details on 6 epidemiological studies since 2004 on diet soda (mainly
aspartame) correlations, as well as 13 other mainstream studies on
aspartame toxicity since summer 2005: Murray 2007.11.14
http://rmforall.blogspot.com/2007_11_01_archive.htm
Wednesday, November 14, 2007
http://groups.yahoo.com/group/aspartameNM/message/1490
"Of course, everyone chooses, as a natural priority, to enjoy peace,
joy, and love by helping to find, quickly share, and positively act upon
evidence about healthy and safe food, drink, and environment."
Rich Murray, MA Room For All
rmforall@...
505-501-2298 1943 Otowi Road, Santa Fe, New Mexico 87505
http://RMForAll.blogspot.com new primary archive
http://groups.yahoo.com/group/aspartameNM/messages
group with 112 members, 1,490 posts in a public,
searchable archive
http://rmforall.blogspot.com/2007_09_01_archive.htm
Saturday, September 15, 2007
http://groups.yahoo.com/group/aspartameNM/message/1472
bias, omissions, incuriosity = opportunity, aspartame safety evaluation,
Magnuson BA, Burdock GA, Williams GM, 7 more, 2007 Sept, Ajinomoto
funded 98 pages html [$ 32 781888262_content.pdf]: Murray 2007.09.15
////////////////////////////////////////////////////////////
[ This layman review gives detailed access to the gist of six
epidemiological studies since 2004, two in 2007, that show correlations
of diet soda (largely aspartame) with health issues.
Probably studies of the correlations at the top 0.1 to 1.0 % level of
use over periods of years by people in vulnerable groups are needed.
http://groups.yahoo.com/group/aspartameNM/message/1141
Nurses Health Study can quickly reveal the extent of aspartame
(methanol, formaldehyde, formic acid) toxicity: Murray 2004.11.21
The Nurses Health Study is a bonanza of information about the health of
probably hundreds of nurses who use 6 or more cans daily of diet soft
drinks -- they have also stored blood and tissue samples from their
immense pool of subjects, over 100,000 for decades.
In total, there are 19 mainstream studies about negative effects with
aspartame since summer, 2005, listed in this review, included many about
the detailed biochemistry involved. ]
////////////////////////////////////////////////////////////
http://RMForAll.blogspot.com September 21, 2007
http://groups.yahoo.com/group/aspartameNM/message/1475
19,000 people, the 4% of users of aspartame who drink average 5 cans
daily, have more problems in NIH AARP study of 474,000 people: Murray
2007.09.21
This is the first good data about the percentage of aspartame users who
use over 3 cans daily, averaging 5 cans daily at 200 mg per 12 oz can
diet soda.
About 4% of 473,984 is 19,000 people, with a peak intake of 17 cans
daily, and average 5 cans daily.
It would be worthwhile to investigate a wide variety of symptoms for the
0.1% of highest level users, about 500 people.
For about 200 million USA aspartame users, this would be 200,000 people.
Table 1 reveals consistent increase in problems from
--------------------- zero to (400 - 600) to (over 600) mg/d
aspartame intake:
% of cohert ---------- 46 -------- 5 -------- 4 %
mean aspartame mg/d --- 0 -------441 ------ 986
16+ education -------- 37 ------- 40 ------- 34 %
diabetes history ------ 3 ------- 22 ------- 26 %
alcohol g/d ---------- 14 ------- 11 ------- 13
never smoke ---------- 36 ------- 31 ------- 29 %
Body Mass Index ------ 26 ------- 29 ------- 29
18.5 - 25 ------------ 42 ------- 21 ------- 19 %
30 - 35 -------------- 13 ------- 23 ------- 26 %
over 35 --------------- 4 ------- 10 ------- 13 %
Physical activity %:
under 3-4/mo --------- 32 ------- 32 ------- 37 %
under 1-2/wk --------- 22 ------- 21 ------- 19 %
over 3-4/wk ---------- 45 ------- 45 ------- 43 %
Calories kcal ----- 1,919 ---- 1,855 ---- 2,044 %
Caffeine mg/d ------- 393 ------ 364 ------ 424
There do seem to be many increases of problems
from the second to third row, as mean aspartame use doubles.
Granted, this is cherry picking the data, selecting interesting patterns.
Correlations alone do not prove any direction of causation.
Nevertheless, it may be of value to study the correlations for
increasing aspartame intake among the 4 % using over 600 mg, the
equivalent of 3 cans 12-oz cans diet soda daily.
The average use for this group is 5 cans daily.
For instance, are a minority of these heavy users displaying the great
majority of the problems that are reflected in the mean for each level
of use, with most users only having little or no increase in problems?
This is a group of about 20,000 people.
"We cannot exclude the possibility that higher aspartame consumption
than that observed in this study may be associated with an elevated risk
of hematopoietic or brain cancers."
http://cebp.aacrjournals.org/cgi/content/full/15/9/1654 free full text
http://cebp.aacrjournals.org/cgi/reprint/15/9/1654 free full text pdf
Cancer Epidemiology Biomarkers & Prevention Vol. 15, 1654-1659,
September 2006
© 2006 American Association for Cancer Research
Consumption of Aspartame-Containing Beverages and Incidence of
Hematopoietic and Brain Malignancies
Unhee Lim 1,
Amy F. Subar 2,
subara@...,
Traci Mouw 1,
Patricia Hartge 1,
Lindsay M. Morton 1,
Rachael Stolzenberg-Solomon 1,
David Campbell 3,
Albert R. Hollenbeck 4
and Arthur Schatzkin 1
1 Division of Cancer Epidemiology and Genetics,
2 Division of Cancer Control and Population Sciences, National Cancer
Institute, NIH, Department of Health and Human Services;
3 Information Management Services, Inc., Rockville, Maryland; and
4 AARP, Washington, District of Columbia
Requests for reprints: Amy Subar,
Division of Cancer Control and Population Sciences,
National Cancer Institute,
6130 Executive Boulevard, EPN 4005, Rockville, MD 20852-7344.
Phone: 301-594-0831; Fax: 301-435-3710. E-mail:
subara@...,
BACKGROUND:
In a few animal experiments, aspartame has been linked to hematopoietic
and brain cancers.
Most animal studies have found no increase in the risk of these or other
cancers.
Data on humans are sparse for either cancer.
Concern lingers regarding this widely used artificial sweetener.
OBJECTIVE:
We investigated prospectively whether aspartame consumption is
associated with the risk of hematopoietic cancers or gliomas (malignant
brain cancer).
METHODS:
We examined 285,079 men and 188,905 women ages 50 to 71 years in the
NIH-AARP Diet and Health Study cohort
Daily aspartame intake was derived from responses to a baseline
self-administered food frequency questionnaire that queried consumption
of four aspartame-containing beverages (soda, fruit drinks, sweetened
iced tea, and aspartame added to hot coffee and tea) during the past year.
Histologically confirmed incident cancers were identified from eight
state cancer registries.
Multivariable-adjusted relative risks (RR) and 95% confidence intervals
(CI) were estimated using Cox proportional hazards regression that
adjusted for age, sex, ethnicity, body mass index, and history of diabetes.
RESULTS:
During over 5 years of follow-up (1995-2000), 1,888 hematopoietic
cancers and 315 malignant gliomas were ascertained.
Higher levels of aspartame intake were not associated with the risk of
overall hematopoietic cancer
(RR for >/=600 mg/d, 0.98; 95% CI, 0.76-1.27),
glioma (RR for >/=400 mg/d, 0.73; 95% CI, 0.46-1.15;
P for inverse linear trend = 0.05),
or their subtypes in men and women.
CONCLUSIONS:
Our findings do not support the hypothesis that aspartame increases
hematopoietic or brain cancer risk. PMID: 16985027
"We cannot exclude the possibility that higher aspartame consumption
than that observed in this study may be associated with an elevated risk
of hematopoietic or brain cancers.
In the laboratory study with positive findings, animals were fed doses
starting from 4 mg up to 5,000 mg per kg body weight.
Significantly elevated lymphomas and leukemias were observed in female
rats fed 20 mg of aspartame and higher (e.g., 1,200 mg for humans
weighing 60 kg or 132 lb; refs. 13, 14).
The reported aspartame intake in our data ranged from 0 to 3,400 mg/d
with sparse numbers in the upper intake categories
(1,200 or 2,000 mg/d, which is equivalent to ~7 to 11 cans of soft
drinks daily) compared with the lowest categories,
and the associations were similarly null in both men and women."
////////////////////////////////////////////////////////////
http://RMForAll.blogspot.com October 12, 2007
http://groups.yahoo.com/group/aspartameNM/message/1479
13,620 seniors using more than 1 can/week artificially sweetened
[aspartame] soft drinks had 8% higher death risk, 1981-2004,
Paganini-Hill A, Kawas CH, Corrada MM, U. Southern Cal., Prev. Med. 2007
April 44(4) 305-10: Murray 2007.10.12
"Individuals who drank more than 1 can/week of artificially sweetened
(but not sugar-sweetened) soft drink (cola and other) had an 8 %
increased risk (95 % CI: 1.01-1.16)."
"The increased death risk with consumption of artificially sweetened,
but not sugar-sweetened, soft drinks suggests an effect of the sweetener
rather than other components of the soft drinks, although residual
confounding remains a possibility."
Prev Med. 2007 Apr; 44(4): 305-10. Epub 2006 Dec 29.
Non-alcoholic beverage and caffeine consumption and mortality: the
Leisure World Cohort Study.
Paganini-Hill A,
annliahi@...,
Kawas CH,
ckawas@...,
Corrada MM.
mcorrada@...,
Department of Preventive Medicine, Keck School of Medicine of the
University of Southern California, CA, USA.
OBJECTIVE:
To examine the effects of non-alcoholic beverage and caffeine
consumption on all-cause mortality in older adults.
METHODS:
The Leisure World Cohort Study is a prospective study of residents of a
California retirement community.
A baseline postal health survey included details on coffee, tea, milk,
soft drink, and chocolate consumption.
Participants were followed for 23 years (1981-2004).
Risk ratios (RRs) of death were calculated using Cox regression for 8644
women and 4980 men (median age at entry, 74 years) and adjusted for age,
gender, and multiple potential confounders.
RESULTS:
Caffeine consumption exhibited a U-shaped mortality curve.
Moderate caffeine consumers had a significantly reduced risk of death
(multivariable-adjusted RR = 0.94, 95 % CI: 0.89, 0.99 for 100-199 mg/day
and RR = 0.90, 95 % CI: 0.85, 0.94 for 200-399 mg/day
compared with those consuming <50 mg/day).
Individuals who drank more than 1 can/week of artificially sweetened
(but not sugar-sweetened) soft drink (cola and other) had an 8 %
increased risk (95 % CI: 1.01-1.16).
Neither milk nor tea had a significant effect on mortality after
multivariable adjustment.
CONCLUSIONS:
Moderate caffeine consumption appeared beneficial in reducing risk of death.
Attenuation in the observed associations between mortality and intake of
tea and milk with adjustment for potential confounders suggests that
such consumption identifies those with other mortality-associated
lifestyle and health risks.
The increased death risk with consumption of artificially sweetened, but
not sugar-sweetened, soft drinks suggests an effect of the sweetener
rather than other components of the soft drinks, although residual
confounding remains a possibility. PMID: 17275898
Age Ageing. 2007 Mar; 36(2): 203-9.
Type of alcohol consumed, changes in intake over time and mortality: the
Leisure World Cohort Study.
Paganini-Hill A, Kawas CH, Corrada MM.
Department of Preventive Medicine,
Keck School of Medicine of University of Southern California, USA.
annliahi@...
BACKGROUND:
modifiable behavioural risk factors including smoking and alcohol
consumption are major contributing or actual causes of mortality.
OBJECTIVE:
to examine the effect of alcohol intake on all-cause mortality in older
adults.
Design and SETTING:
prospective population-based cohort study of residents of a California,
United States retirement community.
SUBJECTS:
8,877 women and 5,101 men (median age, 74 years) who in the early 1980s
completed a postal health survey incluing details on alcohol consumption.
METHODS:
participants were followed for 23 years (1981-2004) including two
follow-up questionnaires (in 1992 and 1998) asking about current alcohol
intake.
Age-adjusted and multivariate-adjusted risk ratios of death and 95 %
confidence intervals were calculated separately for men and women, using
proportional hazard regression.
RESULTS:
of the 8,644 women and 4,980 men with complete information on the
variables of interest and potential confounders,
6,930 women and 4,456 men had died (median age, 87 years).
Both men and women who drank alcohol had decreased mortality compared
with non-drinkers.
Those who drank two or more drinks per day had a 15 % reduced risk of death.
The reduced risk was not limited to one type of alcohol.
Stable drinkers (those who reported drinking both at baseline and
follow-up) had a significantly decreased risk of death compared with
stable non-drinkers.
Those who started drinking at follow-up also had a significantly lower risk.
Women who quit drinking were at increased risk of death.
CONCLUSION:
in elderly men and women, moderate alcohol intake exhibits a beneficial
effect on mortality.
Those who quit may do so for health reasons that affect mortality.
PMID: 17350977
////////////////////////////////////////////////////////////
" Analyses that used food frequency questionnaire data suggested that
intake of over 1 drink per day of either regular or diet soft drinks was
associated with a over 50% higher incidence of metabolic syndrome
compared with intake of under 1 soft drink per week.
" Although the association of high fructose corn syrup intake and
insulin resistance may be a contributory mechanism, 31 in the present
study, both regular and diet soft drinks appeared to pose similar
metabolic hazards,
which suggests that other factors may be operational. "
" The caramel content of both regular and diet drinks may be a potential
source of advanced glycation end products, 5 which may promote insulin
resistance 36 and can be proinflammatory. 37 "
" It is conceivable, though,
that there may be residual confounding caused by lifestyle factors not
adjusted for in the present analyses. "
" As noted above, it is conceivable that residual confounding by
lifestyle/dietary factors not adjusted for may have contributed to the
metabolic risks associated with soft drink intake. "
" The similar metabolic hazard posed by both regular and diet soft
drinks is noteworthy given the lack of calories in the latter; however,
other studies have also reported associations of diet soft drinks with
weight gain in boys 29 and with hypertension in adult women. 7 "
29. Berkey CS, Rockett HRH, Field AE, Gillman MW, Colditz GA.
Sugar-added beverages and adolescent weight change.
Obesity Res. 2004; 12: 778–788.[Abstract/Free Full Text]
7. Winkelmayer WC, Stampfer MJ, Willett WC, Curhan GC.
Habitual caffeine intake and the risk of hypertension in women.
JAMA. 2005; 294: 2330–2335.[Abstract/Free Full Text]
http://circ.ahajournals.org/cgi/content/full/116/5/480 free full text
[ Extracts ]
doi:10.1161/CIRCULATIONAHA.107.689935
CLINICAL PERSPECTIVE
Circulation. 2007; 116: 480-488.
© 2007 American Heart Association, Inc.
Epidemiology
Circulation. 2007 Jul 31; 116(5): 480-8. Epub 2007 Jul 23.
Soft drink consumption and risk of developing cardiometabolic risk
factors and the metabolic syndrome in middle-aged adults in the community.
Ravi Dhingra, MD;
Lisa Sullivan, PhD;
Paul F. Jacques, PhD;
Thomas J. Wang, MD;
Caroline S. Fox, MD;
foxca@...,
James B. Meigs, MD, MPH;
Ralph B. D’Agostino, PhD;
J. Michael Gaziano, MD, MPH;
Ramachandran S. Vasan, MD
vasan@...,
From the National Heart, Lung, and Blood Institute’s Framingham Heart
Study (R.D., T.J.W., C.S.F., R.S.V.), Framingham, Mass;
Massachusetts Veterans Epidemiology Research and Information Center
(R.D., J.M.G.), VA Boston Healthcare System, Boston, Mass;
Division of Aging (R.D., J.M.G.), Brigham and Women’s Hospital, Harvard
Medical School, Boston, Mass; Alice Peck Day Memorial Hospital (R.D.),
Lebanon, NH;
Department of Biostatistics (L.S., R.B.D.), Boston University School of
Public Health, Boston, Mass;
Jean Mayer USDA Human Nutrition Research Center on Aging (P.F.J.), Tufts
University, Boston, Mass; Division of Cardiology (T.J.W.) and Department
of Medicine (J.B.M.), Massachusetts General Hospital, Harvard Medical
School, Boston, Mass;
National Heart, Lung, and Blood Institute (C.S.F.), Bethesda, Md;
Divisions of Preventive Medicine and Cardiovascular Medicine (J.M.G.),
Brigham and Women’s Hospital, Boston, Mass;
and Cardiology Section and the Department of Preventive Medicine and
Epidemiology (R.S.V.), Boston University School of Medicine, Boston, Mass.
Correspondence to Ramachandran S. Vasan, MD, Framingham Heart Study, 73
Mount Wayte Ave, Suite 2, Framingham, MA 01702-5803.
vasan@...,
Received January 12, 2007; accepted May 15, 2007.
BACKGROUND:
Consumption of soft drinks has been linked to obesity in children and
adolescents, but it is unclear whether it increases metabolic risk in
middle-aged individuals.
METHODS AND RESULTS:
We related the incidence of metabolic syndrome and its components to
soft drink consumption in participants in the Framingham Heart Study
(6,039 person-observations, 3,470 in women; mean age 52.9 years) who
were free of baseline metabolic syndrome.
Metabolic syndrome was defined as the presence of over of the following:
waist circumference over 35 inches (women) or over 40 inches (men);
fasting blood glucose over 100 mg/dL;
serum triglycerides over 150 mg/dL;
blood pressure over 135/85 mm Hg;
and high-density lipoprotein cholesterol under 40 mg/dL (men)
or under 50 mg/dL (women).
Multivariable models included adjustments for age, sex, physical
activity, smoking, dietary intake of saturated fat, trans fat, fiber,
magnesium, total calories, and glycemic index.
Cross-sectionally, individuals consuming over 1 soft drink per day had a
higher prevalence of metabolic syndrome
(odds ratio [OR], 1.48; 95 % CI, 1.30 to 1.69)
than those consuming under 1 drink per day.
On follow-up (mean of 4 years), new-onset metabolic syndrome developed
in 765 (18.7 %) of 4095 participants consuming under 1 drink per day and
in 474 (22.6 %) of 2059 persons consuming over 1 soft drink per day.
Consumption of over 1 soft drink per day
was associated with increased odds of developing
metabolic syndrome (OR, 1.44; 95% CI, 1.20 to 1.74),
obesity (OR, 1.31; 95 % CI, 1.02 to 1.68),
increased waist circumference (OR, 1.30; 95 % CI, 1.09 to 1.56),
impaired fasting glucose (OR, 1.25; 95% CI, 1.05 to 1.48),
higher blood pressure (OR, 1.18; 95 % CI, 0.96 to 1.44),
hypertriglyceridemia (OR, 1.25; 95 % CI, 1.04 to 1.51), and
low high-density lipoprotein cholesterol
(OR, 1.32; 95 % CI 1.06 to 1.64).
CONCLUSIONS:
In middle-aged adults, soft drink consumption is associated with a
higher prevalence and incidence of multiple metabolic risk factors.
PMID: 17646581
Key Words: diabetes mellitus • metabolic syndrome • epidemiology •
obesity • risk factors • carbonated beverages
* Introduction
Several reports from the United States and Europe indicate increasing
consumption of soft drinks among children, adolescents, and adults over
the past 3 decades. 1,2
Many clinical studies have linked the rising consumption of soft drinks
to the present epidemic of obesity and diabetes mellitus among children
and adolescents 3–6 and to the development of hypertension in adults. 7
Furthermore, added sweeteners in soft drinks have been linked to an
increase in serum triglycerides levels in some reports 8,9 but not in
others. 10,11
The association of soft drink consumption with obesity and higher
insulin resistance has been attributed to multiple factors, including
greater caloric intake, the high fructose corn syrup content, 12 less
satiety and compensation, and a general effect of consuming refined
carbohydrates (see review by Drewnowski and Bellisle 13).
The aforementioned data raise the possibility that the consumption of
soft drinks can fuel metabolic derangements, including insulin
resistance, that can translate into a greater risk of developing
abdominal obesity, high triglyceride levels, low levels of high-density
lipoprotein cholesterol (HDL-C), elevated blood pressure, and impaired
glucose tolerance; this constellation of metabolic traits has been
collectively referred to as the metabolic syndrome. 14
Higher prevalence of the metabolic syndrome poses greater risk for
cardiovascular disease in the community, 15 although the independent
contribution of this entity to vascular risk beyond its components has
been questioned 16
In the present prospective investigation, we tested the hypothesis that
greater soft drink consumption increases the risk of developing
metabolic risk factors (alone and in combination [metabolic syndrome])
in middle-aged adults in the community.
Additionally, we evaluated whether metabolic risk varied on the basis of
consumption of sugar-sweetened ("regular") versus artificially sweetened
("diet") soft drinks.
* Methods
Study Sample
The Framingham Heart Study began in 1948 with the enrollment of 5,209
participants into the original study cohort. 17
In 1971, children of the original cohort participants and the spouses of
the children were enrolled into the Framingham Offspring Study (n=5,124). 18
Offspring study participants are evaluated approximately every 4 years.
Information on daily consumption of soft drinks was collected via a
physician-administered questionnaire at each study visit from the fourth
(1987–1991) through the sixth (1995–1998) examination cycles.
That examination questionnaire did not elicit information regarding
consumption of regular versus diet soft drinks; however, such
information was available from the self-administered food frequency
questionnaires (FFQ; Willett questionnaire) 19 completed by participants
at the fifth (1992–1995) and sixth examination cycles (see below).
For the present investigation, we selected offspring cohort participants
who attended any 2 consecutive examinations from the fourth through the
seventh (1998–2001) examination cycles.
We excluded participants with missing data on covariates (n = 207) and
those with prevalent cardiovascular disease (n = 926).
After exclusions, a total of 8997 person-observations (4871 in women)
were eligible for the cross-sectional analyses.
For prospective analyses, we excluded individuals with baseline
metabolic syndrome (n = 2897 person-observations; metabolic syndrome as
defined below) and those with any missing metabolic syndrome components
on follow-up (n = 61 person-observations).
The schema for selection of individuals eligible for cross-sectional and
longitudinal analyses is displayed in the Figure.
All participants provided written informed consent, and the protocol for
the study was approved by institutional review board of Boston Medical
Center.
Figure 1185095
Selection of study sample from baseline examinations using the
examination cola questionnaire and from the sample with available FFQ
data (within parentheses, for examinations 5 and 6).
Eligible participants and exclusions are indicated in the Figure.
CVD indicates cardiovascular disease.
Measurement of Covariates
At each Framingham Heart Study examination, participants provided a
medical history and underwent a complete standardized physical
examination that included anthropometry, blood pressure measurements,
and laboratory assessment of vascular risk factors.
Fasting levels of blood glucose, triglycerides, and HDL-C were measured
with standard assays.
Blood pressure was measured by a physician using a mercury
sphygmomanometer and with the participant resting in a seated position
for 5 minutes; the average of 2 readings obtained on the participant’s
left arm constituted the examination blood pressure.
Physical activity was assessed by calculating a "physical activity
index"; participants were asked specific questions regarding how many
hours in a typical day they spent sitting, sleeping, or performing
light-moderate or heavy physical activities. 20
Alcohol intake was assessed by averaging the number of alcoholic
beverages consumed per week.
Participants who reported smoking 1 or more cigarettes per day in the
year before the Framingham Heart Study examination were considered
current smokers.
Assessment of Soft Drink Consumption and Dietary Intake of Other Foods
At the index examinations, participants reported the average number of
12-oz servings of soft drinks (Coke, Pepsi, Sprite, or other carbonated
soft drinks, separately categorized into caffeinated or decaffeinated
drinks) consumed per day in the year preceding the examination.
The responses to the questions were entered as integers (0 or more)
separately for caffeinated and decaffeinated soft drinks.
This questionnaire (referred to as the "examination cola questionnaire")
did not separate nondrinkers from infrequent drinkers (<1 drink per day).
Accordingly, we compared individuals who reported consuming 1, over 1,
or over 2 soft drinks per day with attendees who reported consuming
under 1 soft drink per day (infrequent drinkers and nondrinkers, who
served as the referent).
Intake of regular and diet soft drinks was assessed from FFQs 19 that
were administered at the fifth and sixth examinations.
We also assessed the dietary information on consumption of total
calories, saturated fat, trans fat, fiber, magnesium, and glycemic index
from the FFQ. 19
Because a FFQ was not administered at the fourth examination cycle,
dietary covariate data from the fifth examination cycle were used for
analyses using information from the examination cola questionnaire at
all 3 examinations.
Data from the FFQ were considered valid only if total energy intakes
reported were over 2.51 MJ/d (600 kcal/d) for men and women but under
17.54 MJ/d (4200 kcal/d) for men or under 16.74 MJ/d (4000 kcal/d) for
women and if fewer than 13 food items were left blank.
Each food item was categorized in 9 categories that ranged from never or
under 1 serving per month to over 6 servings per day.
For assessment of saturated fat, trans fat, or dietary fiber, the
nutrient intakes from all specific food items were multiplied by the
frequency of consumption.
The validity of the FFQ has been demonstrated previously. 21
Definition and Components of the Metabolic Syndrome
The metabolic syndrome was considered present if 3 or more of the
following individual components were present 14,22:
waist circumference over 35 inches (88 cm) for
or over 40 inches (102 cm) for men;
fasting blood sugar over 100 mg/dL (5.5 mmol/L) or treatment with oral
hypoglycemic agents or insulin;
blood pressure over 135/85 mm Hg or treatment for hypertension;
serum triglycerides over 150 mg/dL (1.7 mmol/L)
or treatment for hypertriglyceridemia (with niacin or fibrates);
and HDL-C under 40 mg/dL (1.03 mmol/L) in men
or under 50 mg/dL (1.3 mmol/L) in women.
Statistical Analyses
Age- and sex-adjusted baseline characteristics of the participant groups
defined according to the number of soft drinks consumed in 1 day
(under 1, 1, or over 2 per day) were compared by multiple linear and
multiple logistic regression analysis for continuous and categorical
characteristics, respectively.
Data on consumption of soft drinks at each of the 3 eligible baseline
examinations (examination cola questionnaire) were used for this purpose.
Tests for trend in baseline characteristics across soft drink
consumption categories were performed with multiple regression.
We also assessed the baseline characteristics after excluding
participants with prevalent metabolic syndrome at baseline
examinations (sample used for incidence analyses; see below).
Soft Drink Consumption and Prevalence of the Metabolic Syndrome
We used data from examinations 4, 5, and 6 (examination cola
questionnaire) and generalized estimating equations to compare the
prevalence of metabolic syndrome in participants who consumed over 1
soft drink per day with those who consumed under 1 soft drink per day
(referent).
Each participant could contribute up to 3 person-examinations of data
for analysis.
We also evaluated a dose response by comparing individuals
who consumed 1 soft drink per day and those who consumed over 2 soft
drinks per day with the referent group.
We constructed multivariable models in hierarchical fashion with
adjustment for age and sex (model I)
and for age, sex, physical activity index, smoking, dietary consumption
of saturated fat, trans fat, fiber, magnesium, total calories, and
glycemic index (model II).
We used soft drink consumption data from FFQs at examinations 5 and 6,
which yielded a smaller sample (Figure), to relate the prevalence of
metabolic syndrome across the following categories of intake of regular
versus diet soft drinks using generalized estimating equations:
(1) under 1 diet or regular soft drink per week (referent),
(2) 1 to 6 diet soft drinks per week,
(3) over 1 diet soft drink per day,
(4) 1 to 6 regular soft drinks per week,
(5) 1 to 6 regular or diet soft drinks per week,
and (6) over 1 regular soft drink per day.
Individuals reporting consumption of both diet and regular soft drinks
over 1/d (n = 16) were grouped into the last category empirically.
We evaluated the 2 sets of models (I and II) noted above.
Soft Drink Consumption and Incidence of the Metabolic Syndrome
To assess the relations of soft drink consumption to the incidence of
metabolic syndrome, we excluded participants with prevalent metabolic
syndrome at each of examination cycles 4, 5, and 6 (n = 2,897
person-observations).
Then, we used pooled logistic regression analyses
by combining each 4-year follow-up period of observations to relate the
number of soft drinks consumed per day (examination cola questionnaire)
to the incidence of metabolic syndrome (from examination cycles 4 to 5,
5 to 6, and 6 to 7).23
The eligible participants were free of metabolic syndrome
at each baseline examination,
and in this setting, pooled logistic regression has been shown to
provide risk estimates similar to time-dependent Cox models.24
We compared the consumption of soft drinks over 1 per day with
infrequent drinkers (under 1 per day; referent) and also
tested for a dose response by comparing groups consuming 1 and over 2
soft drinks per day with the referent group.
We evaluated 2 sets of models
(covariates as in models I and II above),
which paralleled the analyses of prevalence of metabolic syndrome.
Consumption of soft drinks varies with age and by sex.25
It has also been suggested that the effects of soft drinks and
carbohydrates on metabolic traits may vary according to age, sex,26
and baseline body weight.27
Therefore, we assessed for effect modification by age (modeled
as a continuous variable), sex, and body mass index
(under 30 versus over 30 kg/m2) by incorporating appropriate interaction
terms in the multivariable models.
We repeated analyses with additionally adjustment
for alcohol consumption and baseline levels of systolic and diastolic
blood pressure, blood glucose, serum triglycerides, and HDL-C.
These models were constructed to account for baseline levels of
metabolic traits.
Additionally, we repeated analyses to examine the association
between consumption of caffeinated and decaffeinated soft drinks,
considered separately, and incidence of the metabolic syndrome.
Because individuals with diabetes mellitus are a particularly high-risk
group for developing metabolic abnormalities, we also repeated our
analyses after excluding those with prevalent diabetes mellitus at baseline.
To compare the risk of new-onset metabolic syndrome according to the
type of soft drink consumed (regular versus diet),
we used data from the FFQs at examinations 5 and 6
and evaluated the incidence of the metabolic syndrome across categories
of soft drinks consumed.
The 6 categories of regular and diet soft drinks were those noted above
(for the analyses of the prevalence of metabolic syndrome),
and 2 sets of models were evaluated
(models I and II, as described above).
Incidence of Individual Components of Metabolic Syndrome
We used multivariable logistic regression to evaluate the relations of
soft drink consumption to the incidence of each individual component of
metabolic syndrome using data from the examination cola questionnaire.
We excluded participants who had the specific metabolic trait prevalent
at baseline; for example, we excluded individuals with blood glucose
over 100 mg/dL (5.5 mmol/L) from the "at-risk" group for analysis that
examined the incidence of impaired fasting glucose.
Thus, we examined the incidence of increased waist circumference,
impaired fasting glucose, high blood pressure, hypertriglyceridemia, and
low HDL-C (all defined as above) according to the number of soft drinks
consumed per day.
We evaluated 2 sets of models (I and II, as noted above) and compared
the risk of developing metabolic traits associated with consumption of
over 1 soft drinks per day
with that in infrequent drinkers (under 1 soft drinks per day).
We also evaluated for a dose response as detailed above.
We did not perform analyses of development of individual metabolic
syndrome components in relation to regular versus diet soft drink intake
using the FFQ data at examinations 5 and 6 because the grouping of
incident events into 6 categories resulted in modest numbers of events
in each category.
All analyses were performed with SAS software version 9.0 (SAS
Institute, Cary, NC). A 2-sided probability value of under 0.05 was
considered statistically significant.
The authors had full access to and take full responsibility for the
integrity of the data. All authors have read and agree to the manuscript
as written.
Results
The baseline characteristics of participants according to the categories
of soft drinks consumed per day are presented in Table 1.
Approximately 35 % of the participants reported consuming over 1 soft
drink per day in response to the examination cola questionnaire
(data based on all 3 examinations).
In comparison, only 22 % of participants reported intake of at least 1
soft drink (diet or regular) per day in response to the FFQ (data
available for examinations 5 and 6 only).
The lower proportion reporting daily intake on the FFQ may be related to
the greater number of options available to indicate soft drink intake;
participants drinking 1 to 6 soft drinks per week (also 22 % on the FFQ)
may have rounded their responses on the examination cola questionnaire
to the nearest integer.
View this table:
TABLE 1. Baseline Characteristics of Participants According to
Soft Drink Consumption (n = 8997)
In age- and sex-adjusted models, the prevalence of obesity (assessed
both by body mass index and by waist circumference), high blood
pressure, glucose intolerance, low HDL-C, and hypertriglyceridemia was
significantly higher in those who consumed a greater number of soft
drinks per day.
Serum total cholesterol, low-density lipoprotein cholesterol, physical
activity index, and alcohol consumption did not vary across categories
of soft drinks consumed.
Similar trends were obtained when we excluded individuals with prevalent
metabolic syndrome (Data Supplement, Table I).
Prevalence of the Metabolic Syndrome
There was a 48 % higher adjusted prevalence of metabolic syndrome among
those who consumed 1 or more soft drinks per day relative to individuals
with infrequent soft drink consumption (Table 2).
We observed a rising prevalence of metabolic syndrome across categories
of 1 and over 2 soft drinks per day
In parallel analyses with the data from the FFQ (Table 2), participants
who consumed over 1 diet or regular soft drink per day had nearly a
1.8-fold adjusted prevalence of metabolic syndrome compared with
infrequent drinkers (under 1 per week).
TABLE 2. Cross-Sectional Relationships of Soft Drink Consumption With
Prevalence of Metabolic Syndrome
Incidence of the Metabolic Syndrome
Individuals who consumed at least 1 soft drink per day had a 44 % higher
adjusted risk (95 % CI, 20 % to 74 %) of developing metabolic syndrome
compared with infrequent drinkers in multivariable-adjusted analyses
(Table 3).
There was no effect modification by age, body mass index, or sex
(interaction terms were not statistically significant).
After additional adjustment for baseline levels of covariates (blood
sugar, systolic and diastolic blood pressure, triglycerides, and HDL-C)
and alcohol consumption in our models, the association of consumption of
over 1 soft drink per day with incidence of metabolic syndrome remained
robust (odds ratio [OR], 1.44; 95 % CI, 1.19 to 1.74).
Further exclusion of individuals with diabetes mellitus at baseline (n =
138) attenuated the association (OR for over 1 soft drink per day, 1.16;
95% CI 1.00 to 1.34).
After stratification of analyses by caffeinated versus decaffeinated
drinks, results were consistent with the primary analyses; consumption
of over 1 soft drink per day was associated with incident metabolic
syndrome for both types of beverages (Data Supplement, Table II).
TABLE 3. Multiple Logistic Regression Examining Soft Drink Consumption
and Incidence of Metabolic Syndrome (n = 6154)
In analyses with FFQ data (Table 3), intake of at least 1 regular or
diet soft drink per day was associated with a over 50 % higher incidence
of metabolic syndrome than among those who drank under 1 soft drink per
week, although the association was borderline significant for intake of
over 1 regular soft drink per day ( P = 0.07 ).
We also observed a graded increase in the risk of metabolic syndrome
from those who were consuming 1 to 6 diet or regular soft drinks per
week to those who drank over 1 soft drinks per day (diet or regular).
Incidence of Individual Components of the Metabolic Syndrome
Compared with infrequent drinkers, individuals who consumed over 1 soft
drink per day had a 25 % to 32 % higher adjusted risk of incidence of
each individual metabolic trait (Table 4), with the exception of
development of high blood pressure, for which there was a borderline
significant 18 % higher adjusted odds ( P = 0.10).
TABLE 4. Multiple Logistic Regression Analysis Examining the Relations
of Incidence of Individual Components of Metabolic Syndrome According to
Soft Drink Consumption (Data From All 3 Examinations [4, 5, and 6])
Discussion
In the present study, we observed a significantly higher prevalence of
metabolic syndrome among middle-aged adults who consumed over 1 soft
drink per day.
This association was consistent for intake of both regular and diet soft
drinks.
Our prospective analyses corroborated the cross-sectional findings;
we observed an increase in the incidence of metabolic syndrome among
adults consuming at least 1 soft drink per day, regardless of whether it
was of the regular or diet type.
Additionally, consumption of soft drinks daily was associated with a
higher incidence of each metabolic syndrome component.
The present study extends results from prior studies that reported that
a greater intake of soft drinks is associated with increased prevalence
of metabolic syndrome, 28 higher risk of obesity, 4–6 high blood
pressure, 7 and diabetes mellitus. 5
The similar metabolic hazard posed by both regular and diet soft drinks
is noteworthy given the lack of calories in the latter; however, other
studies have also reported associations of diet soft drinks with weight
gain in boys 29 and with hypertension in adult women. 7
Mechanisms
There are several mechanisms that can explain the higher risk of
metabolic abnormalities associated with greater consumption of soft drinks.
These can be broadly grouped under physiological effects, dietary
behavior, and the economics of food choice. 13
There are several physiological effects of soft drinks that may pose an
adverse metabolic risk.
Larger consumption of added nutritive sweeteners such as high fructose
corn syrup (the primary sweetener in soft drinks) can lead to weight
gain, increased insulin resistance, 30,31 a lowering of HDL-C, 32 and an
increase in triglyceride levels. 27
Typically, in the United States, the high fructose corn syrup added to
the beverages contains about 55 % fructose. 30,31
Although the association of high fructose corn syrup intake and insulin
resistance may be a contributory mechanism, 31 in the present study,
both regular and diet soft drinks appeared to pose similar metabolic
hazards, which suggests that other factors may be operational.
Consumption of liquids is associated with a lesser degree of dietary
compensation (the adjustment in energy intake made in subsequent meals
in response to food intake).
Some investigators believe that intake of sugar-sweetened beverages
induces less compensation than intake of artificially sweetened soft
drinks, 33 but others disagree. 34
The high sweetness of diet or regular soft drinks may lead to
conditioning for a greater preference for intake of sweetened items, 35
although this explanation also has been questioned by some experts. 13
The caramel content of both regular and diet drinks may be a potential
source of advanced glycation end products, 5 which may promote insulin
resistance 36 and can be proinflammatory. 37
Dietary behavior among individuals consuming soft drinks may account in
part for the clustering of metabolic risk factors in these people. 13
Individuals with greater intake of soft drinks also have a dietary
pattern characterized by greater intake of calories and saturated and
trans fats, lower consumption of fiber 38 and dairy products, 39 and a
sedentary life. 40
These observations were corroborated by the our findings of increased
consumption of saturated and trans fat, lower consumption of dietary
fiber, and higher rates of smoking in those with greater intake of soft
drinks.
Nonetheless, in the present investigation, we adjusted for saturated fat
and trans fat intake, dietary fiber consumption, smoking, and physical
activity in multivariable analyses and still observed a significant
association of soft drink consumption with the risk of developing
metabolic syndrome and its component traits.
It is conceivable, though, that there may be residual confounding caused
by lifestyle factors not adjusted for in the present analyses.
Last, it has been suggested that the obesity-promoting effects of soft
drinks may be related in part to their costs, with less expensive drinks
being associated with greater hazard by virtue of their preferential
selection for economic reasons. 13
The present investigation could not explore this explanation.
Strengths and Limitations
The strengths of the present study include the large community-based
sample of men and women and the adjustments for potential confounders;
however, several limitations merit comment.
We chose to use the modified definition of metabolic syndrome
recommended by the National Cholesterol Education Program 14 and did not
use other criteria for the syndrome (such as those suggested by the
World Health Organization 41 or the European panel).
Researchers have found high correlation between these guidelines. 42
Given the observational nature of the present study, we cannot infer
that the observed associations are causal.
As noted above, it is conceivable that residual confounding by
lifestyle/dietary factors not adjusted for may have contributed to the
metabolic risks associated with soft drink intake.
Finally, participants in the present study were all white Americans,
which may limit the generalizability of our results to nonwhites.
Conclusions
In our large community-based sample of middle-aged adults, soft drink
consumption was associated with higher risk of developing adverse
metabolic traits and the metabolic syndrome.
The present observational data raise the possibility that public health
policy measures to limit the rising consumption of soft drinks in the
community may be associated with a lowering of the burden of metabolic
risk factors in adults.
Acknowledgments
Sources of Funding
This work was supported through National Institutes of Health/National
Heart, Lung, and Blood Institute contracts N01-HC-25195, 1R01HL67288,
and 2K24HL04334 (Dr Vasan) and K23HL74077 (Dr Wang) and by a career
development award from the American Diabetes Association (Dr Meigs).
Disclosures
None.
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p 488 CLINICAL PERSPECTIVE
Consumption of soft drinks among children, adolescents, and middle-aged
adults has risen in the United States and Europe during the past 3 decades.
Prior studies have shown a higher prevalence of obesity and diabetes
mellitus in children who consume more soft drinks, although these
associations are less clear for adults.
We evaluated the relations of metabolic syndrome and its components to
soft drink consumption in Framingham participants.
Cross-sectionally, individuals consuming at least 1 soft drink per day
had about 50 % higher prevalence of the metabolic syndrome than those
consuming under 1 drink per day.
During a follow-up period of about 4 years, consumption of over 1 soft
drink per day was associated with a higher incidence of metabolic
syndrome and a higher incidence of each of its components, ie, obesity,
increased waist circumference, impaired fasting glucose, higher blood
pressure, hypertriglyceridemia, and low high-density lipoprotein
cholesterol.
Analyses that used food frequency questionnaire data suggested that
intake of over 1 drink per day of either regular or diet soft drinks was
associated with a over 50% higher incidence of metabolic syndrome
compared with intake of under 1 soft drink per week.
We conclude that consumption of more than 1 soft drink per day is
associated with a higher prevalence and incidence of multiple metabolic
risk factors in middle-aged adults.
Our observational data raise the possibility that public health measures
to limit consumption of soft drinks may be associated with a lowering of
the burden of cardiometabolic risk factors in adults.
Footnotes
The online-only Data Supplement, consisting of tables, is available with
this article at
http://circ.ahajournals.org/cgi/content/full/CIRCULATIONAHA.107.689935/DC1.
Guest Editor for this article was Gregory L. Burke, MD, MSc.
[ Dr. Gregory L. Burke is Professor and Chair of the Department of
Public Health Sciences at the Wake Forest University School of Medicine.
His research interests include epidemiology and cardiovascular disease,
atherosclerosis and subclinical CVD, measurement issues in epidemiology,
clinical trials of chronic disease prevention, women's health,
translation of scientific data for physicians and the general public,
and alternative strategies for chronic disease prevention. Dr. Burke
received his M.D. from the University of Iowa in 1981.
Departments of Public Health Sciences, Pathology, and Obstetrics and
Gynecology, Wake Forest University School of Medicine,
and Lyndhurst Gynecology Associates, Winston-Salem, NC 27157, USA.
gburke@..., ]
Find additional patient-related information at:
http://www.americanheart.org/presenter.jhtml?identifier=3050553
Related Article:
Issue Highlights
Circulation 2007 116: 457. [Full Text]
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////////////////////////////////////////////////////////////
" When studying individual classes of caffeinated beverages, habitual
coffee consumption was not associated with increased risk of hypertension.
By contrast, consumption of cola beverages was associated with an
increased risk of hypertension, independent of whether it was sugared or
diet cola (P for trend <.001).
Conclusion
No linear association between caffeine consumption and incident
hypertension was found.
Even though habitual coffee consumption was not associated with an
increased risk of hypertension, consumption of sugared or diet cola was
associated with it.
Further research to elucidate the role of cola beverages in hypertension
is warranted. "
" The findings were consistent between the cohorts and were present
across types of soda beverages:
both sugared cola and diet cola beverages were associated with an
increased risk of hypertension (Table 5 and Table 6).
Hence, we speculate that it is not caffeine but perhaps some other
compound contained in soda-type soft drinks that may be responsible for
the increased risk in hypertension.
If these associations are causal, they may have considerable impact on
public health. "
" Finally, an examination of the possible associations between
caffeinated cola beverages and the risk of hypertension
showed that
sugared caffeinated cola (NHS I, P for trend = .03; NHS II, P for trend
<.001) (Table 5)
and diet caffeinated cola (NHS I, P for trend = .02; NHS II, P for
trend <.001) (Table 6)
were positively associated with hypertension in both cohorts. "
" Table 6. Age-Adjusted and Multivariate Relative Risks for Incident
Hypertension According to Frequency of Diet Cola Intake
Glasses or Cans of Diet Cola per Day
under 1 ------- 1 ----------- 2-3 ------- 4 and more --- P for Trend
Nurses’ Health Study I (1990-2002) 53,175 nurses, ages 44-69 in 1990
No. of cases of Incident Hypertension
17,268 ------- 1,154 ---------- 662 --------- 130
% 100 ---------- 6.7 ---------- 3.8 -------- 0.75
#% 32.5 -------- 2.2 ---------- 1.3 -------- 0.25 #% of 53,175
Person-years
479,890 ----- 30,579 --------17,316 ------- 3,173
% 100 -----------6.4 ---------- 3.6 -------- 0.66
Age-adjusted relative risk (95% CI)
1.00 -- 1.16(1.10-1.24)-- 1.23(1.13-1.33)-- 1.37(1.15-1.62)-- under .001
Multivariate relative risk (95% CI)*
1.00 -- 1.07(1.00-1.13) -- 1.06(0.98-1.15) -- 1.16(0.97-1.37)------ .02
Nurses’ Health Study II (1991-2003) 87,369 nurses, ages 27-44 in 1991
No. of cases of Incident Hypertension
10,192 -------- 1,452 -------- 1,358 --------- 449
% 100 ---------- 14.3 ----------- 13.3 --------- 4.4
#% 11.7 --------- 1.7 ------------ 1.6 --------- 0.51 #% of 87,369
Person-years
713,971 ----- 91,144 ------- 77,398 ------- 21,265
% 100 --------- 12.8 --------- 10.8 ---------- 3.0
Age-adjusted relative risk (95% CI)
1.00 -- 1.16(1.10-1.23) -- 1.33(1.26-1.41) -- 1.63(1.49-1.80) under .001
Multivariate relative risk (95% CI)*
1.00 -- 1.05(0.99-1.11) -- 1.09(1.03-1.15) -- 1.19(1.08-1.32) under .001
Abbreviation: CI, confidence interval.
*Adjusted for age, body mass index, intake of alcohol, family history of
hypertension, oral contraceptive use (in Nurses’Health Study II only),
physical activity, and smoking status, as well as the other classes of
beverage. "
http://jama.ama-assn.org/cgi/reprint/294/18/2330?ijkey=ff7fa86b688f2c2e23d9b6185\
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JAMA Vol. 294 No. 18, November 9, 2005
Online Features
Original Contribution
Habitual Caffeine Intake and the Risk of Hypertension in Women
Wolfgang C. Winkelmayer, MD, ScD;
wwinkelmayer@...,
Meir J. Stampfer, MD, DrPH;
stampfer@...,
Walter C. Willett, MD, DrPH;
walter.willett@...,
Gary C. Curhan, MD, ScD
gary.curhan@...,
JAMA. 2005; 294: 2330-2335.
Context
Caffeine acutely increases blood pressure, but the association between
habitual consumption of caffeinated beverages and incident hypertension
is uncertain.
Objective
To examine the association between caffeine intake and incident
hypertension in women.
Design, Setting, and Participants
Prospective cohort study conducted in the Nurses’ Health Studies
(NHSs) I and II of 155,594 US women free from physician-diagnosed
hypertension followed up over 12 years
(1990-1991 to 2002-2003 questionnaires).
Caffeine intake and possible confounders were ascertained from regularly
administered questionnaires.
We also tested the associations with types of caffeinated beverages.
Main Outcome Measure
Incident physician-diagnosed hypertension.
Results
During follow-up, 19.541 incident cases of physician-diagnosed
hypertension were reported in NHS I and 13,536 in NHS II.
In both cohorts, no linear association between caffeine consumption and
risk of incident hypertension was observed after multivariate adjustment
(NHS I, P for trend = .29; NHS II, P for trend = .53).
Using categorical analysis, an inverse U-shaped association between
caffeine consumption and incident hypertension was found.
Compared with participants in the lowest quintile of caffeine
consumption, those in the third quintile had a 13 % and 12 % increased
risk of hypertension, respectively (95 % confidence interval in NHS I, 8
% - 18 %; in NHS II, 6 % - 18 %).
When studying individual classes of caffeinated beverages, habitual
coffee consumption was not associated with increased risk of hypertension.
By contrast, consumption of cola beverages was associated with an
increased risk of hypertension, independent of whether it was sugared or
diet cola (P for trend <.001).
Conclusion
No linear association between caffeine consumption and incident
hypertension was found.
Even though habitual coffee consumption was not associated with an
increased risk of hypertension, consumption of sugared or diet cola was
associated with it.
Further research to elucidate the role of cola beverages in hypertension
is warranted.
Author Affiliations:
Division of Pharmacoepidemiology and Pharmacoeconomics (Dr Winkelmayer),
Renal Division (Drs Winkelmayer and Curhan),
and Channing Laboratory (Drs Stampfer, Willett, and Curhan),
Department of Medicine, Brigham and Women’s Hospital, Harvard Medical
School,
and Departments of Epidemiology (Drs Stampfer, Willett, and Curhan) and
Nutrition (Drs Stampfer and Willett), Harvard School of Public Health,
Boston, Mass.
RELATED LETTERS
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Caffeine and Incident Hypertension in Women
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Caffeine and Incident Hypertension in Women—Reply
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JAMA. 2006;295:2137.
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INTRODUCTION
Approximately 50 million people in the United States have hypertension,
and the prevalence is increasing. 1
Hypertension is a major risk factor for coronary heart disease, stroke,
and congestive heart failure. 2-3
Therefore, even small reductions in the prevalence of hypertension could
have a potentially large public health and financial impact.
Much clinical lore about the possible association between caffeine
intake and the risk of hypertension is available.
Short-term studies have demonstrated that caffeine intake acutely
increases blood pressure, but over time, attenuation of this effect does
occur. 4
Experimental studies have shown that caffeine can raise plasma levels of
several stress hormones, such as epinephrine, norepinephrine, 5-6 and
cortisol, all of which can lead to an increase in blood pressure. 6-7
However, these experiments have been limited to relatively short
periods of observation, typically less than 1 week; information on a
more sustained neuroendocrine response to regular exposure to caffeine
is not available.
A long-term effect of caffeine intake on the risk of developing
hypertension would be of substantial public health importance given the
widespread consumption of beverages containing caffeine, but currently,
studies of this association are scarce.
A recent longitudinal study in 1,017 men found a positive association
between coffee consumption and blood pressure and incident hypertension
in unadjusted analyses. 8
Although the association with blood pressure level was significant in
multivariate analyses, a nonsignificant 40 % increase in the risk of
incident diagnosis of hypertension (95 % confidence interval [CI], –6 %
to 109 %) for 3 to 4 cups per day and a 43 % increase (95% CI, –6 % to
118 %) for 5 or more cups per day vs no coffee consumption was found.
No published studies to date of the association between caffeine intake
and the risk of hypertension in women are available.
To prospectively elucidate whether caffeine intake or consumption of
certain caffeine-containing beverages is associated with the risk of
incident hypertension in women, we examined these questions in 2 large
cohort studies of women, the Nurses’ Health Studies (NHSs) I and II.
METHODS
Study Populations
The NHS I cohort was assembled in 1976 when 121,700 female registered
nurses, aged 30 to 55 years, completed and returned a mailed
questionnaire. 9
Follow-up questionnaires have been mailed every 2 years to update
information on health-related behaviors and medical events.
The NHS II began in 1989, when 116,671 female registered nurses, aged 25
to 42 years, completed and returned a mailed questionnaire.
Questionnaires have been mailed every 2 years to update exposure
information and diagnosis of new diseases.
The follow-up for both cohorts exceeds 90 %.
In this analysis, all participants who had not been diagnosed with
hypertension before the return of the 1990 NHS I or 1991 NHS II
questionnaires were included.
This study was approved by the institutional review board at Brigham and
Women’s Hospital, Boston, Mass.
Receipt of each questionnaire implies participant’s consent.
Dietary Assessment
Food frequency questionnaires were used to measure dietary intake and
were completed in 1990, 1994, and 1998 for NHS I and 1991, 1995, and
1999 for NHS II.
Participants were asked about their usual intake of foods and beverages
during the past year.
The response options for specified serving sizes were the following:
never or less than once per month;
1 to 3 times per month;
1 per week;
2 to 4 per week;
5 to 6 per week;
1 per day;
2 to 3 per day;
4 to 5 per day;
and 6 or more per day.
The relevant beverages included on the questionnaire were the following:
low-calorie cola (eg, Diet Coke or Diet Pepsi with caffeine),
regular cola (eg, Coke, Pepsi,
or other cola beverages with sugar),
tea with caffeine, tea without caffeine,
coffee with caffeine, and decaffeinated coffee.
Total caffeine intake was calculated primarily using US Department of
Agriculture food composition sources.
In these calculations, it was assumed that the content of caffeine was
137 mg per cup of coffee, 47 mg per cup of tea, 46 mg per can or bottle
of cola beverage, and 7 mg per serving of chocolate candy. 10
This method of measuring coffee intake was shown to be valid in both the
NHS I cohort and a similar cohort study of male health professionals. 11-13
Assessment of Other Variables
Data on height and family history of hypertension were collected at
baseline in both cohorts.
Information on weight was updated every 4 years.
Using each participant’s updated weight, body mass index was calculated
by dividing the weight in kilograms by height in meters squared.
Also, an updated variable for weight difference between baseline and the
time of respective follow-up questionnaire was generated.
Information on oral contraceptive use in the NHS II cohort also was
updated every 4 years.
The same semiquantitative food frequency questionnaires were used to
determine intake of alcohol, sodium, potassium, magnesium, calcium, and
phosphorus. 14
Physical activity was assessed in NHS I (1988, 1992, and 1996) and NHS
II (1989, 1993, and 1997) cohorts; energy expenditure was expressed in
metabolic equivalent tasks. 15
In addition, the frequency of analgesic drug use (aspirin, nonsteroidal
anti-inflammatory drugs, and acetaminophen) was ascertained. 16-17
Outcome Definition
The baseline and biennial follow-up questionnaires inquired about
physician-diagnosed hypertension and the year of diagnosis.
Self-reported diagnosis of hypertension was found to be reliable in the
NHS I cohort. 18
In a subset of women who reported hypertension, review of medical
records confirmed a documented systolic and diastolic blood pressure,
respectively, higher than 140 mm Hg and 90 mm Hg in 100 % and higher
than 160 mm Hg and 95 mm Hg in 77 % of participants.
Additionally, self-reported hypertension was predictive of subsequent
cardiovascular events. 18
A study participant was considered to have a history of hypertension if
she reported a diagnosis of high blood pressure on any questionnaire up
to and including the 1990 questionnaire in NHS I and the 1991
questionnaire in NHS II, and therefore was excluded from the study.
Among the remaining women in each cohort, incident cases were included
as those who first reported hypertension on any of the subsequent
biennial questionnaires and whose date of diagnosis was after the return
of the 1990 NHS I or the 1991 NHS II questionnaire.
This method recently has been used in a study of folate intake and the
risk of hypertension in women. 19
Statistical Methods
The time of observation was between return of the 1990 NHS I and 1991
NHS II and the 2002 NHS I and 2003 NHS II questionnaires.
Participants who did not return the baseline questionnaires for this
study were allowed to contribute person-time for later time intervals,
provided that they had not been diagnosed with hypertension prior to
return of the respective questionnaire.
Participants were censored after being diagnosed with hypertension or at
the time of death.
Each cohort was analyzed separately.
Age-adjusted Cox proportional hazards regression models were used to
estimate relative risks and 95% CIs.
In addition, multivariate models were constructed that adjusted for
other known risk factors of the study outcome:
age (continuous), body mass index (continuous), alcohol use (6
categories), physical activity (quintiles of metabolic equivalent
tasks), smoking status (current, past, or never), family history of
hypertension (yes/no), and current oral contraceptive use (yes/no; only
in NHS II).
In additional analyses, we ensured that sodium, magnesium, calcium,
potassium, and phosphorus intake (quintiles) did not confound the
estimates from these multivariate models.
All variables were updated to reflect the most recent value provided by
the participants on the biennial questionnaires.
Participants with missing data were assigned to a missing category for
that specific time period.
We determined P values for trend for each of the exposures of interest
by using the median for each category.
Level of significance for P values for trend was <.05.
Also the interaction between caffeine intake and the other variables was
tested.
We used SAS version 8.2 for UNIX statistical software package
(SAS Institute Inc, Cary, NC).
RESULTS
In NHS I, 53,175 women had not been diagnosed with hypertension at
baseline in 1990.
Another 7,916 participants who did not respond to the 1990 questionnaire
but who did respond to a later questionnaire disclosing that they
previously had not been diagnosed with hypertension allowed them to
contribute person-time from that point in time.
Over the 12 years (539,388 person-years of follow-up), 19,541 incident
cases of physician-diagnosed hypertension were reported.
In NHS II, 94,503 participants who were free of hypertension (87,369 in
1991 and an additional 7,134 at a later point in time) were included in
the analyses of younger women.
During 909,199 person-years of observation, 13,536 participants
responded that they were diagnosed with hypertension by a physician.
Participant characteristics by quintile of caffeine intake are presented
in Table 1.
In both cohorts, mean caffeine consumption ranged from less than 20 mg/d
in the lowest quintile to approximately 600 mg/d in the highest quintile.
Caffeine intake was correlated positively with alcohol consumption and
smoking status
r = 0.12, P < .001 for NHS I; r = 0.23, P < .001 for NHS II),
whereas all other relevant characteristics did not differ
materially across quintiles of caffeine consumption.
Table 1. Baseline Characteristics of Cohort by Quintile of Caffeine
Intake in Nurses’ Health Study I (N = 53,175)
and Nurses’ Health Study II (N = 87,369)*
Age-adjusted analyses demonstrated an inverse U-shaped relation between
caffeine intake and the incidence of hypertension in both cohorts.
Compared with participants in the lowest quintile of caffeine
consumption, the risk of incident hypertension was increased by 14 % (95
% CI, 9 % -19 % for NHS I) and 15 % (95 % CI, 9 % - 21 % for NHS II) for
those in the third quintile, whereas those in the highest quintile were
not at an increased risk of hypertension (Table 2).
Multivariate adjustment did not materially change these findings (Table 2).
Table 2. Age-Adjusted and Multivariate Relative Risks for Incident
Hypertension According to Quintile of Caffeine Intake
To further examine this inverse U-shaped association, the frequency of
use of different caffeine-containing beverages in relation to the risk
of incident hypertension was evaluated.
In multivariate models including beverage type, rather than actual
caffeine intake, no association between frequency of intake of
caffeinated coffee and incident hypertension was observed in either cohort.
Compared with NHS I participants drinking less than 1 cup per day of
caffeinated coffee, the relative risks were
1.06 (95% CI, 1.01-1.10) for those consuming 1 cup per day,
1.00 (95% CI, 0.97-1.04) for those drinking 2 to 3 cups per day,
0.93 (95% CI, 0.88-0.99) for those drinking 4 to 5 cups per day,
and 0.88 (95% CI, 0.80-0.98) for those drinking 6 or more cups per day
(Table 3).
The trend for the NHS I cohort was marginally significant for
an inverse association between coffee intake and the risk of
hypertension (Table 3; P for trend = .02).
The findings in the NHS II cohort were practically identical (P for
trend = .03).
The results for intake of decaffeinated coffee also were similar to the
data for caffeinated coffee intake (data not shown);
the trend suggested an inverse association of risk of hypertension in
the NHS I cohort (P for trend = .08)
but not in the NHS II cohort (P for trend = .67).
Table 3. Age-Adjusted and Multivariate Relative Risks for Incident
Hypertension According to Frequency of Coffee Intake
An association between caffeinated tea intake and incident hypertension
in the NHS I cohort (Table 4; P for trend = .79) was not found.
However, in the cohort of younger women in NHS II, a moderate increase
in risk of hypertension (P for trend = .01; Table 4) was detected.
Table 4. Age-Adjusted and Multivariate Relative Risks for Incident
Hypertension According to Frequency of Caffeinated Tea Intake
Finally, an examination of the possible associations between caffeinated
cola beverages and the risk of hypertension
showed that
sugared caffeinated cola (NHS I, P for trend = .03; NHS II, P for trend
<.001) (Table 5)
and diet caffeinated cola (NHS I, P for trend = .02; NHS II, P for
trend <.001) (Table 6)
were positively associated with hypertension in both cohorts.
Table 5. Age-Adjusted and Multivariate Relative Risks for Incident
Hypertension According to Frequency of Sugared Cola Intake
Table 6. Age-Adjusted and Multivariate Relative Risks for Incident
Hypertension According to Frequency of Diet Cola Intake
Glasses or Cans of Diet Cola per Day
under 1 ------- 1 ----------- 2-3 ------- 4 and more --- P for Trend
Nurses’ Health Study I (1990-2002) 53,175 nurses, ages 44-69 in 1990
No. of cases of Incident Hypertension
17,268 ------- 1,154 ---------- 662 --------- 130
% 100 ---------- 6.7 ---------- 3.8 -------- 0.75
#% 32.5 -------- 2.2 ---------- 1.3 -------- 0.25 #% of 53,175
Person-years
479,890 ----- 30,579 --------17,316 ------- 3,173
% 100 -----------6.4 ---------- 3.6 -------- 0.66
Age-adjusted relative risk (95% CI)
1.00 -- 1.16(1.10-1.24)-- 1.23(1.13-1.33)-- 1.37(1.15-1.62)-- under .001
Multivariate relative risk (95% CI)*
1.00 -- 1.07(1.00-1.13) -- 1.06(0.98-1.15) -- 1.16(0.97-1.37)------ .02
Nurses’ Health Study II (1991-2003) 87,369 nurses, ages 27-44 in 1991
No. of cases of Incident Hypertension
10,192 -------- 1,452 -------- 1,358 --------- 449
% 100 ---------- 14.3 ----------- 13.3 --------- 4.4
#% 11.7 --------- 1.7 ------------ 1.6 --------- 0.51 #% of 87,369
Person-years
713,971 ----- 91,144 ------- 77,398 ------- 21,265
% 100 --------- 12.8 --------- 10.8 ---------- 3.0
Age-adjusted relative risk (95% CI)
1.00 -- 1.16(1.10-1.23) -- 1.33(1.26-1.41) -- 1.63(1.49-1.80) under .001
Multivariate relative risk (95% CI)*
1.00 -- 1.05(0.99-1.11) -- 1.09(1.03-1.15) -- 1.19(1.08-1.32) under .001
Abbreviation: CI, confidence interval.
*Adjusted for age, body mass index, intake of alcohol, family history of
hypertension, oral contraceptive use (in Nurses’Health Study II only),
physical activity, and smoking status, as well as the other classes of
beverage.
Additional analyses adjusting for intake of sodium, magnesium,
potassium, phosphorus, and calcium or analgesic drug use did not change
the results materially for the caffeine intake or specific beverage
intake analyses. When testing the robustness of the results, such as by
limiting the analysis to those women who reported having had a routine
physical examination during the time interval or by using baseline body
mass index and updated change in weight rather than updated body mass
index, the results were virtually unchanged (data not shown).
COMMENT
In this prospective study of the association between caffeine intake and
the risk of physician-diagnosed hypertension in 2 large cohorts of
women, we found a modest inverse U-shaped association between caffeine
intake and hypertension in both cohorts.
The magnitude of the highest multivariate relative risk was 1.13 in NHS
I and 1.12 in NHS II.
To better understand this nonlinear relation between caffeine intake and
the risk of hypertension, we evaluated the individual associations of
several caffeine-containing beverages.
Neither caffeinated nor decaffeinated coffee demonstrated a positive
association with incident hypertension in either cohort.
The results for consumption of caffeinated tea were inconclusive:
although no association was observed in the NHS I cohort, a positive
trend was shown in the NHS II cohort.
By contrast, we found a highly significant association between cola
intake (sugared or low-calorie cola) and incident hypertension that was
consistent across the cohorts.
To our knowledge, this study is the first to prospectively evaluate the
putative effect of caffeine consumption on the long-term risk of
hypertension in women.
The speculation that coffee may cause hypertension was supported by
several small experiments over short periods of observation ( under 80
days). 20
If the short-term effects of caffeine on blood pressure persist, then
habitual coffee drinking might contribute to an excess risk of hypertension.
Such an effect would be of great public health importance given the
widespread use of coffee and other caffeinated beverages.
In this study with more than 1.4 million person-years of follow-up, the
relevant exposures and outcomes have been found valid and accurate,
11-13,18 and coffee intake was updated to reflect changes in individual
behavior.
We found strong evidence to refute speculation that coffee consumption
is associated with an increased risk of hypertension in women.
The associations found between caffeinated tea consumption and the risk
of hypertension differed between the 2 cohorts.
In the NHS I cohort, no association was found; however, in the NHS II
cohort, a significant positive trend was observed.
A recent study conducted among 711 men and 796 women in Taiwan found a
strong inverse association between both frequency and duration of tea
intake and hypertension. 21
Since the types of tea (green or oolong) consumed in that study are
likely different from those consumed in our study of US women, the
comparability of the findings from these 2 studies appears uncertain.
In both NHS cohorts we found a positive association between frequency of
caffeinated soft drink consumption and the risk of hypertension.
The findings were consistent between the cohorts and were present across
types of soda beverages: both sugared cola and diet cola beverages were
associated with an increased risk of hypertension (Table 5 and Table 6).
Hence, we speculate that it is not caffeine but perhaps some other
compound contained in soda-type soft drinks that may be responsible for
the increased risk in hypertension.
If these associations are causal, they may have considerable impact on
public health.
Recent studies have found an effect of the intake of cola beverages on
insulin resistance in a rat model 22; in humans, the intake of cola
beverages was associated with an increased risk of diabetes in the NHS
II cohort. 23
These studies have attributed these associations to the glycemic load of
corn syrup, which is used as sweetener in these beverages, and the
caramel coloring, which is rich in advanced glycation end products.
Further studies on the possible mechanisms underlying these associations
clearly are needed.
We acknowledge the limitations of this study.
We cannot rule out that individuals susceptible to adverse effects of
caffeinated coffee intake on their blood pressure in the past may have
reduced their consumption of beverages containing caffeine.
Patients were asked about the frequency of their food intake, but no
information was available on the daily timing of such ingestion.
We did not directly measure the participants’ blood pressure and the
diagnosis of hypertension was self-reported.
Nonetheless, self-reported blood pressure has been validated and
demonstrated to be a strong predictor of actual values. 18
Furthermore, we do not know whether these findings are generalizable
beyond populations of predominantly white women.
We also cannot exclude the possibility that the associations found are
residually confounded.
Lastly, no statement can be made on the effect of coffee intake on the
control of blood pressure among individuals already diagnosed with
hypertension.
In conclusion, consumption of coffee in women does not appear to
increase the risk of developing hypertension.
Whether caffeinated soft drinks are causally related to the risk of
hypertension and its underlying mechanism will require further study.
AUTHOR INFORMATION
Corresponding Author: Wolfgang C. Winkelmayer, MD, ScD, Division of
Pharmacoepidemiology and Pharmacoeconomics and Renal Division, Brigham
and Women’s Hospital, 1620 Tremont St, Suite 3030, Boston, MA 02120
wwinkelmayer@...,
Author Contributions: Dr Winkelmayer had full access to all of the data
in the study and takes responsibility for the integrity of the data and
the accuracy of the data analysis.
Study concept and design: Winkelmayer, Willett, Curhan.
Acquisition of data: Stampfer, Willett, Curhan.
Analysis and interpretation of data: Winkelmayer, Stampfer, Willett, Curhan.
Drafting of the manuscript: Winkelmayer.
Critical revision of the manuscript for important intellectual content:
Winkelmayer, Stampfer, Willett, Curhan.
Statistical analysis: Winkelmayer, Willett, Curhan.
Obtained funding: Willett, Curhan.
Administrative, technical, or material support: Stampfer, Willett, Curhan.
Study supervision: Curhan.
Financial Disclosures: None reported.
Funding/Support:
This study was funded by National Institutes of Health grants DK52866,
DK66574, CA87969, and CA050385.
Dr Winkelmayer is a 2004 T. Franklin Williams Scholar in Geriatric
Nephrology and a recipient of the American Society of
Nephrology-ASP-Junior Development Award in Geriatric Nephrology, jointly
sponsored by the Atlantic Philanthropies, the American Society of
Nephrology, the John A. Hartford Foundation, and the Association of
Subspecialty Professors.
He is also supported by an American Heart Association Scientist
Development grant (0535232N).
Role of the Sponsors:
None of the funding organizations had any role in the design and conduct
of the study; collection, management, analysis, and interpretation of
the data; or preparation, review, or approval of the manuscript.
Author Affiliations
Division of Pharmacoepidemiology and Pharmacoeconomics (Dr Winkelmayer),
Renal Division (Drs Winkelmayer and Curhan), and Channing Laboratory
(Drs Stampfer, Willett, and Curhan), Department of Medicine, Brigham and
Women’s Hospital, Harvard Medical School, and Departments of
Epidemiology (Drs Stampfer, Willett, and Curhan) and Nutrition (Drs
Stampfer and Willett), Harvard School of Public Health, Boston, Mass.
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" Our finding that drinking diet soda during the past year was
associated with weight gain in boys was somewhat unexpected,
particularly because a recent double-blind randomized controlled trial
of overweight adults compared the effects of sucrose and artificial
sweeteners (primarily in beverages) and demonstrated weight loss for the
latter group (71).
However, our overweight boys were drinking nearly 3 times as much diet
soda as our normal weight boys (0.3 vs. 0.1 servings/d)
but similar quantities of regular soda (0.6 servings/d for both groups).
The correlation between year-to-year changes in regular soda and diet
soda intakes was null (r = –0.008; p = 0.70),
suggesting that heavier boys were not substituting diet soda for sugared
soda.
We believe that this explains the diet soda estimate for boys and,
furthermore, illustrates the importance of confirming findings using
blinded randomized trials whenever feasible and ethical.
Because our estimates became considerably smaller after adjusting for
total energy intake, calories probably explain the associations between
beverages and weight gain.
However, we cannot differentiate between calories in the beverages and
calories in foods typically consumed alongside certain beverages (46),
or whether drinking beverages leads to higher subsequent energy intakes
because compensation for energy consumed in liquid form is less complete
than energy consumed in solid form (45) (47) (48). "
Obesity Research 12: 778-788 (2004)
© 2004 The North American Association for the Study of Obesity
Original Research
Sugar-Added Beverages and Adolescent Weight Change
Catherine S. Berkey*,
catherine.berkey@...,
Helaine R.H. Rockett*,
helaine.rockett@...,
Alison E. Field{dagger},
alison.field@...,
Matthew W. Gillman{d dagger},¶
matthew_gillman@...,
and Graham A. Colditz*,§
colditzg@...,
* Channing Laboratory, Department of Medicine, Brigham and Women’s
Hospital and Harvard Medical School, Boston, Massachusetts; Departments
of {ddagger} Nutrition and § Epidemiology, Harvard School of Public
Health, Boston, Massachusetts;
{dagger} Division of Adolescent Medicine, Department of Medicine and
Department of Psychiatry, Children’s Hospital Boston and Harvard Medical
School, Boston, Massachusetts;
and ¶ Department of Ambulatory Care and Prevention, Harvard Medical
School and Harvard Pilgrim Health Care, Boston, Massachusetts.
Address correspondence to Catherine S. Berkey, Channing Laboratory,
181 Longwood Avenue, Boston, MA 02115. E-mail:
catherine.berkey@...,
Abstract
TOP
Abstract
Introduction
Research Methods and Procedures
Discussion
References
Objective:
The increase in consumption of sugar-added beverages over recent decades
may be partly responsible for the obesity epidemic among U.S. adolescents.
Our aim was to evaluate the relationship between BMI changes and intakes
of sugar-added beverages, milk, fruit juices, and diet soda.
Research Methods and Procedures:
Our prospective cohort study included >10,000 boys and girls
participating in the U.S. Growing Up Today Study.
The participants were 9 to 14 years old in 1996 and completed
questionnaires in 1996, 1997, and 1998.
We analyzed change in BMI (kilograms per meter squared) over two 1-year
periods among children who completed annual food frequency
questionnaires assessing typical past year intakes.
We studied beverage intakes during the year corresponding to each BMI
change, and in separate models, we studied 1-year changes in beverage
intakes, adjusting for prior year intakes.
Models included all beverages simultaneously; further models adjusted
for total energy intake.
Results:
Consumption of sugar-added beverages was associated with small BMI gains
during the corresponding year
(boys: +0.03 kg/m2 per daily serving, p = 0.04;
girls: +0.02 kg/m2, p = 0.096).
In models not assuming a linear dose-response trend,
girls who drank 1 serving/d of sugar-added beverages gained more weight
(+0.068, p = 0.02) than girls drinking none,
as did girls drinking 2 servings/d (+0.09, p = 0.06)
or 3+ servings/d (+0.08, p = 0.06).
Analyses of year-to-year change in beverage intakes provided generally
similar findings;
boys who increased consumption of sugar-added beverages from the prior
year experienced weight gain (+0.04 kg/m2 per additional daily serving,
p = 0.01).
Children who increased intakes by 2 or more servings/d from the prior
year gained weight (boys: +0.14, p = 0.01; girls +0.10, p = 0.046).
Further adjusting our models for total energy intake substantially
reduced the estimated effects, which were no longer significant.
Discussion:
Consumption of sugar-added beverages may contribute to weight gain among
adolescents, probably due to their contribution to total energy intake,
because adjustment for calories greatly attenuated the estimated
associations.
Key Words: soda • juice • milk • energy intake • longitudinal
Introduction
Large increases over recent decades in the prevalence of childhood
obesity are well documented (1) (2) (3) (4) (5) (6) (7) (8) (9), as are
the associated health and social consequences of obesity (3) (7) (8) (9)
(10) (11) (12) (13) (14) (15) (16) (17) (18) (19) (20) (21) (22) (23)
(24) (25) (26) (27) (28) (29).
This rapid increase in obesity prevalence implicates environmental
factors (27) (30) (31) (32) (33) (34) (35).
During this time, physical activity among adolescents has declined,
whereas time spent in sedentary activities such as watching
television or videos and playing computer games has increased (5) (6).
Furthermore, in nationally representative samples of U.S. adolescents,
intakes of sugar-added beverages, including soda, have increased (36)
(37) (38) (39).
Higher soft drink intakes are associated with lower milk and fruit juice
intakes and with higher total energy intakes (40).
The largest source of added sugars in the U.S. diet is nondiet soft
drinks (37)
One cross-sectional study of dietary intakes (41) has reported similar
soda and fruit drink intakes for obese vs. nonobese adolescents,
whereas another has found a positive correlation between measures of
adiposity in adolescents and soft drink intakes (42).
However, two other studies have suggested an inverse association between
adiposity and intake of sugars (43) (44).
Ludwig et al. (45) published the first longitudinal analysis of
sugar-added beverage intakes and body weight changes.
They followed 548 ethnically diverse 11- and 12-year-old children in
Boston-area public schools for 19 months and found positive associations
among sugar-sweetened beverage intakes, weight change, and incident
obesity.
Whether the critical factor is the sugar, the calories, or behaviors
related to beverage consumption is unknown.
Aside from the calories within each beverage, some foods may frequently
accompany certain beverages (46), and drinking beverages may also lead
to higher subsequent energy intakes because compensation for energy
consumed in liquid form is less complete, due to lower satiety, than
energy consumed in solid form (45) (47) (48).
Furthermore, sugar-added beverages may encourage additional energy
intake because of their high glycemic index (49).
It may be informative to further consider all beverages simultaneously
and to study children from a broader age range and with longer follow-up.
Using data from the Growing Up Today Study, an ongoing prospective
cohort study of children from all over the U.S., we analyzed the
relationship between intakes of beverages
(milk, sugar-added beverages, fruit juices, and diet soda)
and changes over time in BMI.
Research Methods and Procedures
Study Population
Established in the fall of 1996, the Growing Up Today Study consists of
16,771 children, residing in 50 states, who are offspring of Nurses’
Health Study II (NHSII)1 participants (50).
The study is described in detail elsewhere (51).
These children were ages 9 through 14 years old in 1996.
In 1997 and 1998, we sent subjects follow-up questionnaires to update
all information.
Response rates to at least one of these follow-ups were 92.5% for girls
and 87.7% for boys.
Measurements
BMI.
Children self-reported their height and weight annually on our
questionnaire, which provided specific measuring instructions but
suggested that they ask someone for help.
Because their mothers are nurses who biennially self-report their own
height and weight as part of NHSII, assistance was available to each of
them.
A previous study reported high validity for self-reported heights and
weights for children 12 to 16 years old (52).
We assessed adiposity by computing BMI = weight/(height)2 squared
(kilograms per meter squared).
The International Obesity Task Force supports the use of BMI to assess
fatness in children and adolescents (53).
Childhood BMI is related to other measures of adiposity that were not
feasible to include in our large self-report study (54) (55).
A recent study (56) supports the validity of BMI computed from
self-reported height and weight, with a correlation of 0.92 between BMI
computed from measured values and self-reports by youth in grades 7
through 12.
Before computing BMIs, we excluded any height that was >3 SD beyond the
gender-age-specific mean height (0.46% of heights excluded) and any
1-year height change which declined by >1 inch or increased by >3 SD
above the mean change (1.65% excluded).
We further excluded any BMI < 12.0 kg/m2 as a biological lower limit
(clinical opinion, MWG) and BMI > 3 SD above or below the
gender-age-specific mean [log(BMI) scale] (0.87% excluded).
We then estimated our outcome, annual change in adiposity, by
BMI1997 – BMI1996 and BMI1998 – BMI1997, dividing each by the exact time
interval between the pair of measurements.
Because all these represented realistic 1-year changes in weight, there
were no further BMI exclusions: 6,871 girls and 5,321 boys provided BMI
change data.
Unfortunately, no validation studies of change in BMI, derived from
self-reported data, have been conducted.
We grouped children, based on their BMI at the earlier year of each
1-year time interval, using the Centers for Disease Control and
Prevention (CDC) gender- and age-specific percentiles for BMI (57) .
Children above the 85th percentile were at risk of overweight (85th to
95th percentile), and those above the 95th percentile were overweight
(57).
Similarly, we grouped together those below the 10th percentile
for BMI.
For simplicity, we refer to all children whose BMI exceeded the
85th percentile as "overweight," those below the 10th percentile as
"very lean," and those between the 10th and 85th percentiles as "normal
weight."
The CDC standards were also used to assign age-specific z scores to BMIs.
Beverages.
Members of our research group designed a self-administered
semiquantitative food frequency questionnaire (FFQ), specifically for
older children and adolescents, which is inexpensive and simple to
administer to large populations (58).
This FFQ for youth has been shown to be valid and reproducible on
children 9 through 18 years old (58) (59);
the mean correlation for nutrients from the FFQ compared with three
24-hour recalls was r = 0.54,
which is comparable with the performance of a similar adult FFQ.
The youth FFQ included questions regarding frequency of intake of 132
specific food items over the past year.
Beverage questions indicated that the serving size was a can, glass,
bottle, or cup (tailored to the particular beverage).
The question about "Hawaiian Punch, lemonade, Koolaid, or other
noncarbonated fruit drink" preceded questions about "orange juice" and
"apple juice and other fruit juices."
For each beverage, we derived typical past year intake (servings per
day) and change in intakes between years.
We also estimated total energy intake (kilocalories per day) and
excluded as implausible intakes <500 or >5000 kcal/d (0.53% excluded).
The beverages we studied were
sugar-added beverages (soda, sweetened iced tea, and noncarbonated fruit
drinks), fruit juices (orange juice and apple/other juices), diet soda,
and milk (white, in a glass or on cereal, and chocolate).
Alcohol and coffee intakes were very low;
therefore, we did not include them.
Physical Activity.
We developed a physical activity questionnaire, specifically for youth,
which asked the participants to recall the typical amount of time spent,
within each season over the past year, in 17 activities and team sports
(outside of gym class); response categories ranged from 0 to 10+ h/wk.
From each child’s responses, we computed his/her typical hours of
weekly physical activity within each season and over the entire year.
Assessments of an earlier nonseasonal version of this instrument found
that estimates of total physical activity were moderately reproducible
and reasonably correlated with cardio-respiratory fitness, thus
providing evidence of validity (60).
Another validation study reported a correlation of r = 0.80 between
survey self-reports and 24-hour recalls in sixth to eighth grade
children (61).
We developed the seasonal version used in this paper to further improve
reliability and validity (62).
Estimates of total physical activity that exceeded 40 h/wk were deemed
implausible and excluded (3.8%).
Inactivity.
Another series of questions was designed to measure weekly hours of
recreational inactivity: "watching TV," "watching videos or VCR," and
"Nintendo/Sega/computer games (not homework)."
For each of these, children reported their usual number of total hours,
separate for weekdays and for weekends, from options ranging from 0 to
31+ hours.
From this information, we computed each child’s typical hours of
recreational inactivity per week.
Gortmaker and colleagues (61) reported moderate reproducibility for
children in grades six to eight for recalled total inactivity from a
similar instrument.
We considered totals exceeding 80 h/wk implausible and excluded them
(0.94%).
Race/Ethnicity.
At baseline, children reported their race/ethnic group by marking all of
six options that applied.
We assigned each child to one of five racial/ethnic groups following
U.S. Census definitions, except that we retained Asians as a separate
group rather than pooled with "other" (1).
Tanner Stage, Menarche, and Age.
Each year, children reported their Tanner maturation stage, a validated
self-rating (63) of sexual maturity that uses five
categories/illustrations for stage of pubic hair development,
and girls reported whether/when their menstrual periods began.
We derived a menstrual history variable having three categories:
premenarche both before and after the 1-year BMI change,
periods that began during the interval,
and postmenarche both years.
We computed each child’s age from dates of birth and questionnaire return.
Statistical Analyses
To assess the potential for selection bias, we compared the baseline
(1996) values of age, BMI, individual beverage intakes, and total energy
intakes of those children who returned surveys in consecutive follow-up
years with those who did not.
The differences were small (see "Results").
All models throughout were fit separately for boys and girls.
Cross-Sectional Analyses.
We reported gender- and age-specific means at baseline for height,
weight, total energy intake, and daily intakes of seven beverages.
A linear regression model related baseline total daily energy intakes to
the intake of each beverage.
Longitudinal Analyses.
To study the effects of beverage intakes during the year of BMI change,
we related the past year typical beverage intakes reported in 1997 to
change in BMI from 1996 to 1997 and intakes reported in 1998 to BMI
change from 1997 to 1998.
Because each child can have two BMI changes,
the assumption of independent observations required by ordinary
regression models was not met, so we used mixed linear regression models
(64) with estimation by SAS proc mixed (65).
We also estimated the effects of 1-year change in beverage intakes (the
difference between intakes in 1996 and 1997 and between 1997 and 1998)
on same-year change in BMI.
The prior year intake (reported in 1996 and 1997) was included as a
covariate in the mixed model.
All models adjusted for race/ethnicity, and to account for increases in
BMI that typically occur during growth and maturation, we included
height growth during the same year, menstrual history, Tanner stage,
prior BMI z score, and nonlinear age trends (30) (66) (67) (68) (69)
(70).
Models also adjusted for activity and inactivity during the year
of BMI change (51) and for milk type (whole/2%/1%/nonfat/soy).
We included total energy intake in further models as a hypothesized
intermediary in the pathway between beverages and weight gain.
Results
These children, whose mothers are all participants in the NHSII (50),
are mostly white (94.7%).
At baseline, 23.2% of the boys and 17.5% of the girls were overweight
(>85th percentile on CDC BMI charts),
whereas 7.2% of the boys and 8.6% of the girls were very lean (<10th
percentile).
Children who did not return surveys in adjacent years (required for
inclusion in our longitudinal analyses) were slightly older (girls by
0.20 years; boys by 0.32 years; both p < 0.05).
At baseline, they drank slightly less milk (girls by 0.18 servings/d;
boys by 0.11 servings/d)
but more sugar-added beverages
(girls by 0.13 servings/d; boys by 0.10 servings/d)
(each age-adjusted p < 0.05).
There were no significant differences at baseline in age-adjusted BMI,
total energy intake, fruit juice intake,
or diet soda intake (each age-adjusted p > 0.05).
Cross-Sectional Results
Older children drank less milk but more orange juice, soda, iced tea,
and punch than younger children (Table 1).
Boys reported higher energy intakes and drank more milk, punch, orange
juice, and soda than did same-age girls.
At baseline, children who drank more milk and less diet soda were leaner,
whereas girls who drank more sugar-added beverages were heavier
(BMI +0.06 kg/m2 higher per serving, p = 0.04).
Table 1. Baseline means for total energy intakes (kilocalories
per day), beverage intakes (servings per day), height (inches), and
weight (pounds) for youth participating in the Growing Up Today Study
To explore whether drinking certain beverages may be linked to higher
total energy intakes, we related daily total energy intake to each of
the beverages separately (Table 2) .
As expected, diet soda intakes were not associated with higher total
energy intakes.
Milk intakes were associated with total energy intakes, with per serving
effects slightly more than the energy provided by the milk, whereas the
per serving effects for sugar-added beverage and fruit juice intakes
were considerably larger than their own energy contents.
Table 2. Cross-sectional association between beverage intakes and
total energy intakes at baseline*
Longitudinal Results
Among children who completed the FFQ all 3 years, mean milk intake
declined significantly each year,
whereas soda intake increased significantly (Figure 1).
Apple juice intake declined for both boys and girls between 1996 and 1997,
diet soda and orange juice intake each increased for girls between 1997
and 1998,
and orange juice intake increased for boys each year (all p < 0.05).
Figure 1. Mean beverage intakes in children from the Growing Up
Today Study who provided dietary data in all 3 years of follow-up. All
year-to-year increases in soda intake and declines in milk intake were
statistically significant.
Beverage Intakes During Year of BMI Change.
We related BMI changes over 1-year periods to beverage intakes during
the same year. The regression coefficients ß (Table 3) represent the
1-year change in BMI (kilograms per meter squared) expected per usual
daily serving of each beverage.
For boys, intakes of sugar-added beverages (ß = 0.03)
and diet soda (ß = 0.12) were significantly associated with weight gains;
there were suggestions (p < 0.06) that intakes of milk (ß = 0.02)
and fruit juices (ß = 0.04) were also associated.
After including total energy intake in the model, the estimated ßs for
all beverages (except diet soda) were nearly one-half their unadjusted
magnitudes and no longer significant (p > 0.31; Table 3 ).
For girls, our analysis suggested (p < 0.10) a linear association
between 1-year weight gain and intakes of milk (ß = 0.02) and
sugar-added beverages (ß = 0.02); the corresponding energy-adjusted
estimates were slightly smaller (all p > 0.15; Table 3 ).
Table 3. Longitudinal analysis of beverage intakes and change in
BMI (kilograms per meter squared) during the same time period*
Figure 2 (far left) presents the association between BMI change and
sugar-added beverages analyzed as a categorical variable (0, 1, 2, or 3+
servings/d) to permit nonlinear trends; all Figure 2 models adjusted for
all covariates except energy intake. A dose-response trend was confirmed
for boys, consistent with the statistically significant per-serving
effect (also shown in Figure 2 ). Girls who reported one (0.5 to 1.5)
daily serving of sugar-added beverages gained significantly more BMI
(0.068 kg/m2, p = 0.02) during the year than those reporting none (0 to
<0.5 servings) (Figure 2 , far left). Girls consuming two (+0.09, p =
0.06) or three+ servings (+0.08, p = 0.06) also gained weight compared
with nondrinkers.
Figure 2. Sugar-added beverages: association between past year
intake (left) or 1-year change in intake (right), and 1-year change in
BMI. Estimates are shown separately for number of servings per day
compared with none and for the per-serving effect (assumes a linear
dose-response trend). Models adjusted for all covariates except energy
intake.
Beverage Change and BMI Change.
For boys, increasing sugar-added beverage intake from one year to the
next was significantly associated
(ß = 0.04 per added daily serving; p = 0.01)
with weight gain (Table 4),
and increasing milk and diet soda intakes
were weakly associated (p < 0.10) with weight gain.
With total energy intake (prior year energy and change in energy) in the
model, estimates for the energy-containing beverages each declined by
>40%, and none remained significant (Table 4).
For girls, increasing intake of sugar-added beverages was weakly
linearly related to weight gain
(ß = 0.03, p = 0.08); energy adjustment diminished the estimated effect
(p = 0.16).
Table 4. Longitudinal analysis of change in beverage intakes and
change in BMI (kilograms per meter squared) over the same year,
adjusting for prior beverage intakes*
Figure 2 (right half) shows that boys who increased their sugar-added
beverage intake by 1 serving/d from the previous year gained more weight
(+0.10 kg/m2, p = 0.02) than boys with unchanged intake, and those who
increased their intake by 2 or more servings/d gained even more (+0.14,
p = 0.01).
Girls (Figure 2 , right) who increased their intake by 1 serving/d over
the previous year gained marginally more BMI (+0.065, p = 0.079)
than girls whose intakes were unchanged, and girls whose intakes
increased by 2 or more servings/d
gained significantly more BMI (+0.10, p = 0.046).
Combining Energy-Containing Beverages.
Because the models in Tables 3 and 4 suggested that any of the beverages
containing calories might contribute to male weight gains, we combined
together these beverages (total servings per day of milk, sugar-added
beverages, and fruit juices).
For boys, this total was associated with weight gain
(ß = +0.03 kg/m2 per daily serving during the year of BMI change, p =
<0.01;
and ß = +0.03 per increase in daily serving from the prior year, p =
<0.01).
For girls, because the ßs for fruit juice were <0 in Tables 3 and 4,
combining fruit juice intakes with milk and sugar-added beverages
did not provide a significant association
(ß = +0.01, p = 0.096 per daily serving during the year of BMI change;
and ß = +0.01, p = 0.13 per increase in daily serving from the prior year).
After energy adjustment, significant effects became smaller by at least
31% and were no longer significant (p > 0.12).
Discussion
Although a previous publication (45) considered whether sugar-added
beverages contribute to weight gain among 11- to 12-year-old children,
we addressed the effects of several types of beverages on children 9 to
17 years old.
Our strongest and most consistent evidence was a linear association
between sugar-added beverage intakes (past year and change from prior
year) and weight gain in boys (both p < 0.05).
The evidence for girls was less compelling but still suggestive (p <
0.10) of a linear association between sugar-added beverages and weight gain.
Girls who drank 1 serving/d during the past year gained more weight than
nondrinkers (p < 0.05).
Both boys and girls who increased their intakes by 2 or more servings/d
from the previous year experienced significant weight gain, as did boys
who increased their intakes by 1 serving/d from the previous year.
However, the magnitudes of these estimated effects were small;
a boy consuming 3 servings/d of sugar-added beverages over
10 years is expected to gain only 0.9 BMI more than if he consumed none.
Our finding that drinking diet soda during the past year was associated
with weight gain in boys was somewhat unexpected,
particularly because a recent double-blind randomized controlled trial
of overweight adults compared the effects of sucrose and artificial
sweeteners (primarily in beverages) and demonstrated weight loss for the
latter group (71).
However, our overweight boys were drinking nearly 3 times as much diet
soda as our normal weight boys (0.3 vs. 0.1 servings/d)
but similar quantities of regular soda (0.6 servings/d for both groups).
The correlation between year-to-year changes in regular soda and diet
soda intakes was null (r = –0.008; p = 0.70),
suggesting that heavier boys were not substituting diet soda for sugared
soda.
We believe that this explains the diet soda estimate for boys and,
furthermore, illustrates the importance of confirming findings using
blinded randomized trials whenever feasible and ethical.
Because our estimates became considerably smaller after adjusting for
total energy intake, calories probably explain the associations between
beverages and weight gain.
However, we cannot differentiate between calories in the beverages and
calories in foods typically consumed alongside certain beverages (46),
or whether drinking beverages leads to higher subsequent energy intakes
because compensation for energy consumed in liquid form is less complete
than energy consumed in solid form (45) (47) (48).
The compensation theory that liquid foods have lower satiety than solid
foods would apply to milk and fruit juice as well as to sugar-added
beverages.
Sugar-added beverages may also encourage further energy intake because
of their high glycemic index (49).
Under any of these possible mechanisms, consumption of sugar-added
beverages encourages higher total energy intakes, which promotes weight
gain, so that adjusting models for energy should diminish the estimated
associations.
The fact that sugar-added beverages lost statistical significance in
energy-adjusted models does not imply that sugar-added beverages are not
responsible for weight gain because of the pathway.
The literature regarding cross-sectional associations between adiposity
and beverage consumption is mixed (41) (42) (43) (44) (72).
Our cross-sectional results indicated that heavier children were
drinking less milk and more diet soda, presumably to lose weight or
prevent further weight gain, although girls who drank sugar-added
beverages tended to be heavier.
In a nationally representative sample of U.S. children, BMI was
positively associated with diet carbonated beverages and, for girls,
negatively associated with milk intakes (70).
We further presented cross-sectional evidence similar to Harnack et al.
(40) that drinking sugar-added beverages was associated with higher
total energy intakes.
The first longitudinal study of sugar-sweetened beverages (45), on 548
ethnically diverse 11- and 12-year-old children in Boston-area public
schools, reported associations between change in beverage intakes from
baseline to 19 months later and BMI change.
Their study differed from ours in that they did not have Tanner Stage
data, their FFQ and report of activity/inactivity related to past month
(ours was past year), and they did not study milk intakes.
Their BMIs were measured rather than self-reported, which may partially
explain why their estimate for a single serving per day increase
[ß = +0.20 kg/m2 over 19 months, not energy adjusted; from their Table 2
(45) ] is larger than our estimate
(over 12 months: boys ß = +0.10 kg/m2, p = 0.02; girls +0.07, p = 0.08).
A major strength of our analysis was the longitudinal design, which
allowed us to study changes over time in beverage intakes and in BMI
while accounting for growth and maturation.
BMI typically goes up from year to year among children in this age
range, and we took these changes into account.
Although our observational study cannot provide conclusive evidence of
causality, our evidence is stronger than that obtainable from
cross-sectional studies.
Baseline differences between children excluded and included in our
longitudinal analyses were small, though the loss of some children with
higher intakes of sugar-added beverages (0.1 more servings/d)
could bias our estimates of those effects.
Because we included all beverages together in our models, we minimized
confounding by other beverage intakes.
However, residual and unmeasured confounding is still possible despite
extensive control for many important covariates.
A major limitation of our study was the necessity of collecting data
(including height, weight, and beverage intakes by FFQ) on youth by
self-report on mailed questionnaires, but with our large geographically
dispersed cohort, alternatives were not feasible.
The impact of random reporting errors should be to bias estimates of
true associations toward the null, possibly explaining why our estimates
were quite small even when statistically significant.
Large soft drink portion sizes complicate the reporting of intakes and
encourage overconsumption (35).
Data collected by 24-hour recalls from 1994 to 1996 (73) showed that
the average soft drink portion size was 19.9 ounces, and differences
were noted among eating locations (home, restaurant, and fast food).
Our FFQ did suggest portion sizes ["soda, not diet (1 can or glass)";
response category "1 can per day"] but did not specify the number of
ounces in a can or glass, so confusion over this may have further
biased our estimates toward the null.
Although we cannot claim that our children of nurses are representative
of U.S. children, the associations among factors within our cohort
should still be internally valid.
Our sugar-added beverage intakes for 11- to 12-year-old children (1.35
servings/d for boys and 1.14 for girls) were similar to those of 11- to
12-year-old children studied by Ludwig et al. (1.22 servings/d) (45).
In 1998, Jacobson summarized the history of soft drink consumption, its
nutritional value, its potential impact on osteoporosis, tooth decay,
heart disease, and kidney stones, and its marketing by the industry,
with recommendations for what should be done (74).
Here, we extend the evidence (45) that sugar-added beverages (which
include soda) may contribute to weight gain.
Even if milk and fruit juice also contribute to weight gain, they have
nutritional benefits, whereas soda provides only calories (74).
The increase in soft drink serving sizes and the increase in energy
intakes provided by soft drinks since 1977 have been documented (73)
(75) (76) , and reversing this trend may help prevent further increases
in obesity prevalence.
For both children and adults, prevention of obesity is critical, and for
weight loss, recommendations include eating a nutritionally balanced,
low-energy diet while increasing energy expenditure through regular
physical exercise (77) (78).
Beverage intakes, including limiting the consumption of soft drinks, are
a potential target for improving diets of adolescents (42) (45) (74) (79).
Data from our cohort suggested that children who reduce intakes of
sugar-added beverages, along with other behavior modifications such as
increasing physical activity and reducing time with TV/videos/computer
games (80) , may prevent excessive weight gains that can lead to obesity.
Acknowledgments
This study was funded by NIH Grant DK46834,
by Boston Obesity Nutrition Research Center Grant P30 DK46200,
by Prevention Research Center Grant U48/CCU115807 from the Centers for
Disease Control and Prevention,
by Research Grant 43-3AEM-0-80074 from the Economic Research Service of
the U.S. Department of Agriculture,
and, in part, by Kellogg’s.
The authors are grateful to Catherine Tomeo Ryan, Karen Corsano, Gary
Chase, and Gideon Aweh for ongoing technical support and to all their
colleagues in the Growing Up Today Study Research Group.
The authors are especially grateful to the children (and their mothers
for encouragement) for careful completion of the questionnaires.
Footnotes
The costs of publication of this article were defrayed, in part, by the
payment of page charges. This article must, therefore, be hereby marked
"advertisement" in accordance with 18 U.S.C. Section 1734 solely to
indicate this fact.
1 Nonstandard abbreviations: NHSII, Nurses’ Health Study II; CDC,
Centers for Disease Control and Prevention; FFQ, food frequency
questionnaire. Back
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