700 ARTICLES Journal of the National Cancer Institute, Vol. 98, No. 10, May 17, 2006
Statin Use and Breast Cancer: Prospective Results From
the Women’s Health Initiative
Jane A. Cauley , Anne McTiernan , Rebecca J. Rodabough , Andrea LaCroix ,
Douglas C. Bauer , Karen L. Margolis , Electra D. Paskett , Mara Z. Vitolins ,
Curt D. Furberg , Rowan T. Chlebowski
For the Women’s Health Initiative Research Group
Background: Despite experimental observations suggesting
that 3-hydroxy-3-methylglutaryl coenzyme A inhibitors
(statins) have antitumor activity, clinical studies have reached
mixed conclusions about the relationship between statin use
and breast cancer risk. Methods: To investigate associations
between potency , duration of use, and type of statin used and
risk of invasive breast cancer, we examined data for 156 351
postmenopausal women who were enrolled in the Women’s
Health Initiative. Information was collected on breast cancer
risk factors and on the use of statins and other lipid-lowering
drugs. Cox proportional hazards regression was used to cal-
culate hazard ratios (HRs) with 95% confi dence intervals
(CIs) . Statistical tests were two-sided. Results: Over an aver-
age follow-up of 6.7 years, 4383 invasive breast cancers were
confi rmed by medical record and pathology report review.
Statins were used by 11 710 (7.5%) of the cohort. Breast can-
cer incidence was 4.09 per 1000 person-years (PY) among
statin users and 4.28 per 1000 PY among nonusers. In multi-
variable models, the hazard ratio of breast cancer among
users of any statin, compared with nonusers, was 0.91 (95%
CI = 0.80 to 1.05, P = .20). There was no trend in risk by dura-
tion of statin use, with HR = 0.80 (95% CI = 0.63 to 1.03) for
<1 year of use, HR = 0.99 (95% CI = 0.80 to 1.23) for 1 – <3 years
of use, and HR = 0.94 (95% CI = 0.75 to 1.18) for ≥ 3 years
of use. Hydrophobic statins (i.e., simvastatin, lovastatin,
and fl uvastatin) were used by 8106 women, and their use was
associated with an 18% lower breast cancer incidence (HR =
0.82, 95% CI = 0.70 to 0.97, P = .02). Use of other statins
(i.e., pravastatin and atorvastatin) or nonstatin lipid-lowering
agents was not associated with breast cancer incidence.
Conclusions: Overall statin use was not associated with inva-
sive breast cancer incidence. Our fi nding that use of hydro-
phobic statins may be associated with lower breast cancer
incidence suggests possible within-class differences that war-
rant further evaluation. [J Natl Cancer Inst 2006;98:700 – 7]
Statins are widely prescribed, effective cholesterol-lowering
drugs. Indeed, atorvastatin and simvastatin were the most com-
monly prescribed drugs in the United States in 2004, with over
70 million prescriptions written for atorvastatin alone ( 1 ) . The
statins are pleiotropic agents, and, after an early study of patients
with coronary heart disease showed a lower than expected inci-
dence of cancers ( 2 ) , preclinical studies were carried out that
have supported the potential anticancer activity of these com-
pounds ( 3 , 4 ) . However, clinical reports on the relationship
between statin use and breast cancer risk have yielded mixed
results, with no association ( 5 , 6 ) and both positive ( 7 , 8 ) and
negative ( 9 ) associations being observed. Prior observational
studies have not evaluated the statin – breast cancer link by statin
potency or category (i.e., hydrophobic versus not). Because
breast cancer is the most frequent cancer in U.S. women, any
link between statin use and breast cancer risk would have major
public health implications.
The objective of the current study was to examine the asso-
ciations between the potency, duration, and type of statin used
and invasive breast cancer risk among women enrolled in the
Women’s Health Initiative (WHI). A secondary objective was to
assess the association between use of other lipid-lowering agents
and breast cancer.
S UBJECTS AND M ETHODS
The WHI includes an observational study ( n = 93 676) and
clinical trials ( n = 68 132) of hormone therapy, dietary modifi ca-
tion, and/or calcium and vitamin D supplementation in postmeno-
pausal women of many races and ethnicities. Recruitment to the
WHI was conducted between October 1, 1993, and December
31, 1998, at 40 clinical centers in the United States. Women were
eligible if they were aged 50 – 79 years, were postmenopausal,
planned to remain in the area where they lived at recruitment , and
had an estimated survival of at least 3 years. Study methods have
been described in detail elsewhere ( 10 ) . This analysis included
Affi liations of authors : Department of Epidemiology, University of Pittsburgh,
Pittsburgh, PA (JAC); Division of Public Health Sciences, Fred Hutchinson
Cancer Research Center, Seattle, WA (AM, RJR, AL); Departments of Medicine
and Epidemiology and Biostatistics, University of California at San Francisco, CA
(DCB); Berman Center for Outcomes and Clinical Research, Hennepin County
Medical Center, Minneapolis, MN (KLM); Comprehensive Cancer Center,
School of Public Health, The Ohio State University, Columbus, OH (EDP);
School of Medicine, Department of Public Health Sciences, Wake Forest
University, Winston-Salem, NC (MZV, CDF); Los Angeles Biomedical Research
Institute at Harbor-UCLA Medical Center, Torrance, CA (RTC).
Correspondence to: Jane A. Cauley, DrPH, University of Pittsburgh, 130
DeSoto St., Crabtree Hall A524, Pittsburgh, PA 15261 (e-mail: email@example.com ).
See “ Notes ” following “ References. ”
© The Author 2006. Published by Oxford University Press. All rights reserved.
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by guest on May 14, 2011
Journal of the National Cancer Institute, Vol. 98, No. 10, May 17, 2006 ARTICLES 701
women enrolled in the observational study and clinical trial com-
ponents of the WHI, excluding those who had previously been
diagnosed with breast cancer or who had used tamoxifen or any
selective estrogen receptor modulator. The fi nal sample included
88 322 women enrolled in the observational study and 68 029
women enrolled in the clinical trials (156 351 women total).
All participants signed informed consent forms. All protocols
and procedures were approved by institutional review boards at
participating institutions. Follow-up for this report is through
February 2004, for a mean ± SD of 6.7 ± 1.5 years.
Participants were asked to bring all current prescription medi-
cations to their fi rst screening interview. Clinic interviewers en-
tered each medication name directly from the containers into the
WHI database, which assigned drug codes using Medispan soft-
ware (First DataBank, Inc., San Bruno, CA). Women reported
duration of use for each current medication. Information on dose
was not recorded. Current medication use was updated at the year
3 clinic visit with identical methods.
Current statin use was defi ned as use of any 3-hydroxy-3-
methylglutaryl coenzyme A (HMG-CoA) reductase inhibitor.
Statins were further classifi ed as hydrophobic (lovastatin, sim-
vastatin, and fl uvastatin) or other (pravastatin and atorvastatin)
and by potency: low (fl uvastatin and lovastatin), medium (prava-
statin), and high (simvastatin and atorvastatin) ( 11 ) . Other lipid-
lowering medications included fi brate, colestipol, probucol,
cholestyramine, niacin, and nicotinic acid.
Breast Cancer Screening and Diagnosis
Medical history was updated annually (in the observational
study) or semiannually (in the clinical trials) by mail and/or tele-
phone questionnaires. For women in the clinical trial components
of the WHI, the frequency of clinical breast examination and
mammography was protocol defi ned (annually for women in the
hormone trials and biennially for women in the dietary trial).
Clinical breast examination and mammography were not proto-
col defi ned for women in the observational study. Data on the
frequency of clinical breast examination and mammography
were collected annually from all participants.
Self-report of breast cancer was locally verifi ed at each clinic
by medical record and pathology report review by centrally
trained WHI physician adjudicators. Central adjudication and
coding of histology, extent of disease, and estrogen receptor (ER)
and progesterone receptor (PR) status (positive or negative per
pathology report) were performed at the Clinical Coordinating
Center using the Surveillance, Epidemiology, and End Results
Program (SEER) coding system ( 12 , 13 ) . Only invasive breast
cancer cases confi rmed by adjudication were included in the
analysis (4383 cases). Information on ER status was available for
3793 invasive breast cancer cases.
Information on all covariates was collected at study entry .
Current and previous use of menopausal hormone therapy and
oral contraceptives were ascertained by interview using a de-
tailed questionnaire that included type, route of administration,
number of pills per day or week, and duration of use for each
hormonal preparation ever taken. Hormone therapy users were
defi ned as those who used estrogen (with or without progestin)
after menopause for at least 3 months.
Baseline questionnaires ascertained information on race or
ethnicity (white, black, Hispanic, American Indian, Asian/Pacifi c
Islander, or unknown), history of physician-diagnosed diabetes
(yes/no), high serum cholesterol level that required treatment
with pills (yes/no), history of myocardial infarction or angina
(yes/no), history of benign breast disease (yes/no), educational
level (<high school, high school diploma/GED, or >high school
diploma/GED), family history of female breast cancer (yes/no),
hysterectomy and oophorectomy status (yes/no), ages at men-
arche ( ≤ 11, 12 – 13, or ≥ 14 years) and fi rst birth (never pregnant, no
term pregnancy, or <20, 20 – 29, or ≥ 30 years), parity (none, 1 – 2,
or ≥ 3), use of nonsteroidal anti-infl ammatory drugs (NSAIDs)
or aspirin (yes/no), current and past smoking status , and time
(minutes per week) spent in mild, moderate, or strenuous physi-
cal activity (none, 10 – <115, or 115 – ≥ 250 minutes/week). Alcohol
consumption (none/past drinker, <1 drink/week, or ≥ 1 drink/
week) and percentage of calories from fat ( ≥ 30% versus <30% of
calories from fat) were estimated from a food- frequency ques-
tionnaire ( 14 ) . Body mass index (BMI) was calculated as weight
in kilograms divided by height in meters squared. The Gail
5-year breast cancer risk estimate was calculated. A woman was
considered at high risk if her Gail score was >1.7% ( 15 ) .
The characteristics of statin users at baseline were compared
with those of nonusers by chi-square or Fisher exact tests (for
categorical variables) or two-sample t tests (for continuous vari-
ables). Incidence rates of breast cancer per 1000 person-years
were calculated according to the use of statins and other lipid-
lowering agents. An a priori plan of analysis specifi ed that we
perform selected subgroup analyses by statin use duration
(<1 year, 1 – <3 years, and ≥ 3 years), potency, and hydrophobic
status. Women who reported using two or more statins were in-
cluded in analyses that compared statin use to none but were ex-
cluded from analyses that examined details of statin use (i.e., by
potency or type). Separate analyses were conducted for women
with ER-positive and ER-negative breast cancer. Hazard ratios
(HRs) for breast cancer among statin users versus nonusers and
95% con fi dence intervals (CIs) were computed from Cox propor-
tional hazards analyses. Tests for the proportional hazards as-
sumptions were conducted by a Cox model that included statin
use and the interaction of statin use with follow-up time and test-
ing for a zero coeffi cient on the interaction term. Results of these
analyses showed that the assumptions were not violated.
All models were adjusted for assignment to active hormone or
placebo in the two WHI hormone trials (estrogen plus progestin
and estrogen alone), assignment to intervention or control in the
dietary modifi cation trial, or enrollment in the observational
study. We also adjusted for prior hormone use at baseline (none,
prior estrogen alone, prior estrogen plus progestin, or prior use of
both estrogen alone and estrogen plus progestin) . These adjust-
ments resulted in what we refer to as the base model. The base
model was further adjusted by age; these age- adjusted base mod-
els included 155 530 women. To control for potential confound-
ing factors, we used multivariable Cox proportional hazards
analyses with a forced-entry approach for variable selection. In
addition to the variables in the age- and base factor – adjusted
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702 ARTICLES Journal of the National Cancer Institute, Vol. 98, No. 10, May 17, 2006
Table 1. Baseline characteristics by statin use *
No statin use Statin use
N % N %
Age at baseline, y
50 – 59
60 – 69
70 – 79
Asian/Pacifi c Islander
None – some high school
High school diploma/GED
>High school diploma/GED
≥ 1 drink/week
Physical activity, min/wk
10 – <115
115 – <250
≥ 30% energy from fat
Body mass index, kg/m 2
25 – <30
Have a current medical
Mammogram in the
last 2 years
Gail risk of breast
Family history of
breast cancer †
Age at menarche, y †
12 – 13
Ever pregnant †
Age at fi rst birth, y †
No term pregnancy
20 – 29
Number of live births †
1 – 2
Benign breast disease
Yes, 1 biopsy
Yes, ≥ 2 biopsies
Hormone therapy use
Current, <5 y
Current, 5 – <10 y
Current, ≥ 10 y
144 64192.5 11 7107.5
116 030 82.910 147 89.3
54 51237.7 522644.6
24 88718.22052 18.6
No statin use Statin use
N % N %
Current HT use by type
(includes HT trial use)
Estrogen plus progestin
Prior oral contraceptive use
General health rating
Estrogen-alone trial only
Estrogen plus progestin
Dietary modifi cation trial only
HT and dietary
modifi cation trials
* P -values from chi-square tests comparing statin users to nonusers are <.001
for all characteristics except as indicated. HT = hormone therapy; NSAID = non-
steroidal anti-infl ammatory drug; WHI = Women’s Health Initiative.
† P values for comparison are not statistically signifi cant.
Table 1 (continued).
models, the multivariable models were adjusted for race and eth-
nicity, BMI, physical activity, current and past smoking, family
history of breast cancer, hysterectomy status, mammogram in the
past 2 years, educational level, ages at menarche and fi rst birth,
parity, alcohol consumption, and percentage of calories from fat.
Multivariable models were based on the 115 683 individuals re-
maining after the exclusion of participants with missing values
for any of the covariates. To evaluate the effects on the results of
change in statin use over time, fi nal models were rerun by enter-
ing statin use as a time-dependent exposure and using updated
information on statin use gathered at the year 3 clinic visit. We
examined the risk for breast cancer by statin use separately in
users of estrogen plus progestin, users of estrogen alone, and
never/past users of hormone therapy.
Comparisons of breast cancer tumor characteristics between
statin users and nonusers were based on chi-square tests, two-
sample t tests, or Brown – Mood tests of medians. All analyses
were conducted using SAS software, version 9.1 (SAS Institute,
Inc., Cary, NC). All statistical tests were two-sided.
In this cohort of 156 351 women, 11 710 (7.5%) were statin
users ( Table 1 ). Women using statins at baseline were older at
enrollment than nonusers (65.6 and 63.0 years, respectively) and
had a higher BMI (mean 28.9 and 27.9 kg/m 2 , respectively).
Statin users were less likely than nonusers to have more than a
high school education, to drink alcohol, to be physically active,
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Journal of the National Cancer Institute, Vol. 98, No. 10, May 17, 2006 ARTICLES 703
to have used hormone therapy, and to report that they obtained
≥ 30% calories from fat. Statin users were more likely than
nonusers to have smoked, to have had a hysterectomy or bilateral
oophorectomy, and to report use of NSAIDs and aspirin. A higher
proportion of statin users reported having had a mammogram in
the past 2 years, although ≥ 80% of both users and nonusers had
had a mammogram in the past 2 years. A higher proportion of
statin users than nonusers were considered at high risk of breast
cancer, i.e., to have a Gail 5-year breast cancer risk of >1.7%. Of
the 2162 women who reported using nonstatin lipid-lowering
agents, 288 women reported also using a statin. Although most of
the absolute differences between statin users and nonusers were
small, many were statistically signifi cant because of the large
number of women in the cohort.
Of the 11 710 statin users, 4591 (39.2%) used a low-potency
statin, 2645 (22.6%) used a medium-potency statin, and 4438
(37.9%) used a high-potency statin ( Table 2 ). A total of 8106
(69.2%) of the women who used statins reported using at least
one hydrophobic statin. A year 3 medication history was avail-
able for 135 772 women (82% of the cohort). Among cohort
members who used statins at baseline, 8274 women (82.5%)
were still using a statin at the year 3 clinic visit; among those who
did not use statins at baseline, 11 583 women newly reported
taking a statin at the year 3 visit.
During a total of 1 041 518 person years (PY) of observa-
tion, 4383 women were diagnosed with invasive breast can-
cer. The incidence of breast cancer was approximately 4.4%
lower among women reporting statin use (4.09 per 1000 PY)
than among nonusers (4.28 per 1000 PY). In the age-adjusted
base model, the relative risk of breast cancer was 8% lower
among statin users than among nonusers (HR = 0.92, 95%
CI = 0.82 to 1.03) ( Table 3 ). In the full multivariable-adjusted
model, breast cancer incidence was approximately 9% lower
in statin users than in non-users (HR = 0.91, 95% CI = 0.80 to
1.05, P = .20).
Examination of the relative risk of breast cancer by duration
of statin use ( Table 3 ) revealed no consistent trend. Short-term
use (<1 year) was associated with a non – statistically signifi cant
20% reduction in invasive breast cancer, whereas use for 1 to 3
years and for more than 3 years was not associated with the risk
of breast cancer.
We also examined breast cancer risk by statin potency and cat-
egory ( Table 3 ). Use of low- and high-potency statins was associated
with non – statistically signifi cant reductions in breast cancer inci-
dence (of 15% and 17%, respectively), but use of medium-potency
statins showed no such association. Use of hydrophobic statins was
associated with a statistically signifi cant 18% reduction in risk of
Table 3. Incidence and hazard ratios of invasive breast cancer by use of statins and other lipid-lowering medications *
Breast cancer cases
per 1000 PY
HR (95% CI) from
age-adjusted base model
HR (95% CI) from
multivariable-adjusted † model
Type of statin
Statin category §
Statin potency ||
Duration of statin use
1 – <3 years
Other lipid-lowering medication
0.92 (0.81 to 1.03)
0.91 (0.80 to 1.05)
0.87 (0.70 to 1.09)
0.84 (0.68 to 1.05)
0.81 (0.57 to 1.15)
1.10 (0.88 to 1.37)
0.99 (0.65 to 1.51)
0.84 (0.65 to 1.09)
0.80 (0.62 to 1.04)
0.80 (0.53 to 1.19)
1.17 (0.92 to 1.49)
1.05 (0.66 to 1.67)
0.85 (0.74 to 0.98)
1.08 (0.89 to 1.31)
0.82 (0.70 to 0.97)
1.14 (0.92 to 1.42)
0.85 (0.71 to 1.03)
1.10 (0.88 to 1.37)
0.87 (0.72 to 1.06)
0.83 (0.66 to 1.03)
1.17 (0.91 to 1.49)
0.85 (0.68 to 1.07)
0.80 (0.65 to 1.00)
0.95 (0.79 to 1.16)
0.99 (0.82 to 1.20)
0.80 (0.63 to 1.03)
0.99 (0.80 to 1.23)
0.94 (0.75 to 1.18)
0.92 (0.71 to 1.19)
0.88 (0.64 to 1.19)
* PY = person-year; HR = hazard ratio; CI = confi dence interval.
† Age-adjusted base model was further adjusted for body mass index, race, smoking, family history of breast cancer, education, hysterectomy, mammogram in the
last 2 years, age at menarche, parity/age at fi rst birth, alcohol use, percentage of calories from fat, physical activity, and nonsteroidal anti-infl ammatory drug use.
‡ Information on specifi c statin use was available for 296 of the 297 statin users with breast cancer.
§ Hydrophobic statins are simvastatin, lovastatin, and fl uvastatin; others are pravastatin and atorvastatin.
|| Low-potency statins are lovastatin and fl uvastatin, the medium-potency statin is pravastatin, and the high-potency statins are simvastatin and atorvastatin.
Table 2. Statin use details for the 11 710 users of any statin
Type of statin used
Two or more statins
Duration of statin use, y
1 – <3
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704 ARTICLES Journal of the National Cancer Institute, Vol. 98, No. 10, May 17, 2006
breast cancer (HR = 0.82, 95% CI = 0.70 to 0.97, P = .02), whereas
use of other statins was not associated with breast cancer incidence
(HR = 1.14, 95% CI = 0.92 to 1.42, P = .24).
To test for possible interactions between statin use and post-
menopausal hormone use, we examined the association between
statin use and breast cancer separately in women who used estro-
gen plus progestin, those who used estrogen alone, and never/past
users of hormones ( Table 4 ). Statin use was not associated with
breast cancer risk among users of estrogen plus progestin or among
never/past hormone users ( Table 4 ). Among users of estrogen
alone, statin use was associated with a non – statistically signifi cant
22% reduction in the risk of breast cancer ( P for the interaction
between hormone use and statin use = .09). The multivariable-
adjusted hazard ratio for ER-positive breast cancer among statin
users compared with non-users was 0.97 (95% CI = 0.83 to 1.13),
and that for ER-negative breast cancer was 0.83 (95% CI = 0.55 to
1.25). The breast cancers in statin users and nonusers were similar
in size, number of positive lymph nodes, SEER stage, histology,
tumor grade, and ER and PR status ( Table 5 ).
Finally, we analyzed breast cancer risk according to the use of
lipid-lowering agents other than statins. The incidence of breast
cancer in users of such agents (4.04 per 10 000 PY) was 5.4%
lower than that in non-users (4.27 per 1000 PY), but the differ-
ence was not statistically signifi cant in the multivariable model
(HR = 0.88, 95% CI = 0.64 to 1.19, P = .41).
The current report is the largest cohort study, to our knowl-
edge, to evaluate statin use and invasive breast cancer in terms
of the number of incident breast cancers. We studied 156 361
women, who were followed for 1 041 518 person-years, and 4383
incident breast cancers. The full multivariable model used in the
analysis adjusted for a comprehensive set of breast cancer risk
factors, including age, race, BMI, family history of breast cancer,
alcohol consumption, physical activity, mammography utiliza-
tion, past and current menopausal hormone therapy, smoking,
percentage of calories from fat, educational level, NSAID use,
and reproductive history. When we considered statins as a class,
we found no association between statin use and breast cancer
risk. Although the relative risk of breast cancer was approxi-
mately 9% lower among statin users than among nonusers, the
difference was not statistically signifi cant. Breast cancer inci-
dence was also not associated with duration of statin use or statin
potency. There was an interaction between statin use and hormone
therapy that was of borderline statistical signifi cance: current
users of both estrogen alone and a statin had a somewhat lower
risk of breast cancer than women who had never used a statin.
This interaction was not observed among users of estrogen plus
progestin, however, and these results also confl ict with results
from the Nurses’ Health Study ( 5 ). Finally, women using hydro-
phobic statins (simvastatin, lovastatin, or fl uvastatin) had an 18%
lower breast cancer incidence than nonstatin users ( P = .02).
Previous reports on statin use and breast cancer risk, which
include both randomized trials of subjects with coronary heart
disease and risk factors for coronary heart disease ( 2 , 8 , 16 – 21 )
and observational studies ( 5 – 7 , 9 , 22 – 27 ) , have provided mixed
results. For example, in two randomized trials of pravastatin in
older people at risk of vascular disease , a nonhydrophobic statin,
more breast cancers were seen in the statin group. In one of these
studies, the Cholesterol and Recurrent Events (CARE) study, one
woman of 291 in the placebo group and 12 women of 291 in the
pravastatin group developed breast cancer ( P = .002) ( 8 ) . Of the
12 breast cancers, however, three occurred in women who previ-
ously had breast cancer and one occurred in a woman who took
pravastatin for only 6 weeks. In the second of these trials of
pravastatin, more breast cancers were diagnosed in women tak-
ing pravastatin than in women taking placebo (HR = 1.65, 95%
CI = 0.78 to 3.49), but the difference was not statistically signifi -
cant ( 19 ) . In contrast, the Long Term Intervention with Prava-
statin in Chronic Disease (LIPID) trial, with 1516 women
randomly assigned to pravastatin or placebo, found no increase
in breast cancer with pravastatin use ( 17 ) . The Heart Protection
study found slightly fewer breast cancers among women ran-
domly assigned to the hydrophobic statin simvastatin than among
those assigned to placebo, but the difference was not statistically
signifi cant ( 18 ) . Other randomized trials of simvastatin ( 21 ) or
pravastatin ( 20 ) found no association with breast cancer risk.
In a recent meta-analysis of 90 056 participants in 14 randomized
trials of statins, statin users did not have an increase in risk of
cancer death or cancer incidence, including breast cancer ( 28 ) .
Among the observational studies, an increase in breast cancer
incidence for statin users has been observed in some case – control
studies ( 4 , 21 ) , especially for short-term users of statins and past
long-term users of hormone therapy ( 7 ) . However, neither a large
case – control study from the General Practice Research Database,
an automated data source containing drug prescription and medical
information on more than 3 million people in the United Kingdom
( 24 ) , nor three other large cohort studies ( 5 , 6 , 26 ) found an associa-
tion between statin use and breast cancer. In a large case – control
study of nearly 1000 women with breast cancer identifi ed from the
Table 4. Incidence of invasive breast cancer by statin and hormone use * at baseline
Current hormone use
No statin use Statin use
HR (95% CI) ‡ P for interaction CasesRate † Cases Rate †
Any E + P
1.09 (0.93 to 1.28)
0.78 (0.60 to 1.02)
0.93 (0.74 to 1.18)
* If a woman had been randomly assigned to the estrogen plus progestin (E + P) group or reported active use of estrogen plus progestin at baseline, she was con-
sidered an estrogen plus progestin user. If a woman was randomly assigned to the active estrogen (E)-alone group or reported estrogen-alone use at baseline, she
was considered in the estrogen-alone group. Women randomly assigned to placebo groups and women who reported past or never use at baseline were considered
† Rate given per 1000 person-years.
‡ HR = hazard ratio; CI = confi dence interval.
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Journal of the National Cancer Institute, Vol. 98, No. 10, May 17, 2006 ARTICLES 705
Cancer Surveillance System, a population-based tumor registry
that serves 13 counties in western Washington State, no overall
association of statins with breast cancer incidence was seen, but
women who had used statins for more than 5 years had an approx-
imately 30% lower breast cancer incidence than never users ( 25 ) .
A statistically signifi cantly decrease in breast cancer incidence in
statin users has been seen in only two ( 9 , 27 ) of eight cohort studies.
However, the number of breast cancers in one of these reports was
small ( 9 ) , and limited information on breast cancer risk factors was
provided in the second report, which was an abstract ( 27 ) .
Our results, taken together with the existing literature, indi-
cate that breast cancer risk is at least not increased in statin users.
Whether or not statin use is associated with reduced breast can-
cer risk is less certain. In the current study, after adjustment for
breast cancer risk factors, statin users had a somewhat lower
breast cancer incidence than nonusers. However, the differences
were statistically signifi cant only in women who reported using
hydrophobic statins. This observation is consistent with a cell
culture study in which only hydrophobic statins (lovastatin, sim-
vastatin, and fl uvastatin) but not a hydrophilic statin (pravastatin)
Table 5. Breast cancer characteristics by statin use
No Statin Use Statin Use
n Mean (SD) or %n Mean (SD) or %
P value *
Tumor size, cm †
No primary mass
Microscopic focus or foci
≤ 0.5 cm
>0.5 – 1 cm
>1 – 2 cm
>2 – 5 cm
Lymph nodes examined
Number of lymph nodes examined †
Number of positive lymph nodes †
Number of positive lymph nodes
1 – 3
Lymph nodes positive
SEER stage §
Ductal and lobular
Estrogen receptor assay
Progesterone receptor assay
1.6 (1.2)189 1.4 (1.4) .33
* P values are from a two-sample t test for continuous variables or from a chi-square test for categorized variables. The fi rst P value for a given characteristic tests
the association with statin use by using only known values of the characteristic. The P value corresponding to the “ missing ” rows tests the association of percent
missing for the given characteristic with statin use.
† Mean (SD) only applies to those with known tumor size or known number of lymph nodes examined or positive.
‡ P value for number of positive lymph nodes based on Brown – Mood test of medians.
§ SEER = Surveillance, Epidemiology, and End Results Program.
by guest on May 14, 2011
706 ARTICLES Journal of the National Cancer Institute, Vol. 98, No. 10, May 17, 2006
had anticancer activity ( 29 ) . Pravastatin may promote the devel-
opment of cancer by causing an induction of mevalonate synthe-
sis in extrahepatic tissues ( 30 ) , an effect that is not observed with
other statins. This increase in mevalonate appears to promote the
growth of breast cancer cells ( 30 ) . In the randomized trials of
statins, an increase in breast cancer was observed only in the two
trials of pravastatin ( 8 , 19 ) . Moreover, in the cohort study that
reported a 72% lower risk of breast cancer among statin users,
the majority of these users (247 of 284) used a hydrophobic statin
( 9 ) . Thus, the inconsistency in previous results may refl ect differ-
ences in the association with specifi c statins.
Considering all other nonstatin lipid-lowering medications
together, we found no statistically signifi cant association be-
tween their use and breast cancer risk. However, because the
multivariable-adjusted relative risk of breast cancer was 12%
lower among users of these other agents than among nonusers
and because one previous cohort study also reported a statisti-
cally signifi cantly lower breast cancer risk among users of other
lipid- lowering agents than among nonusers ( 9 ) , further study of
the infl uence of individual lipid-lowering agents on breast cancer
incidence may be warranted.
Strengths of this study include the prospective design; inclu-
sion of a large, racially diverse sample of well-characterized
women; collection of detailed information on a comprehensive
range of breast cancer risk factors; complete follow-up for
breast cancer outcomes; regular assessment of mammography
use; blinded adjudication of breast cancers via pathology report
review; description of breast cancer histologic characteristics
and hormone receptor status; and the ability to examine associa-
tions by statin category. The limitations of this study include its
observational design. Although we adjusted for many factors
that could confound the association between statin use and
breast cancer, there may be residual confounding by unmea-
sured factors. Indeed, a recent comparison of observational
study and randomized clinical trial results, with respect to fi nd-
ings regarding postmenopausal hormone use and coronary heart
disease, showed that the discrepancy in fi ndings can be substan-
tially explained by confounding ( 31 ). Study limitations also
include the relatively low prevalence of statin use, lack of
information on dose, and limited power to examine long-term
(>5 years) effects.
In conclusion, in this large population of postmenopausal
women with well-characterized breast cancer risk factors, when
all statins were considered together as a class, no statistically sig-
nifi cant association with breast cancer incidence was seen. How-
ever, use of hydrophobic statins was associated with statistically
signifi cantly lower breast cancer incidence, a fi nding that war-
rants further evaluation. Future studies of statins and breast can-
cer should assess associations with individual statins or statin
categories because class differences may exist.
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The Women’s Health Initiative (WHI) program is funded by the National Heart,
Lung and Blood Institute, U.S. Department of Health and Human Services. The
sponsor played a role in the design and analysis of the WHI.
A short list of WHI investigators is as follows:
Program Offi ce: ( National Heart, Lung, and Blood Institute, Bethesda, MD)
Barbara Alving, Jacques Rossouw, Linda Pottern.
Clinical Coordinating Center: (Fred Hutchinson Cancer Research Cen-
ter, Seattle, WA) Ross Prentice, Garnet Anderson, Andrea LaCroix, Charles
L. Kooperberg, Ruth E. Patterson, Anne McTiernan; (Wake Forest University
School of Medicine, Winston-Salem, NC) Sally Shumaker; (Medical Research
Laboratories, Highland Heights, KY) Evan Stein; (University of California at San
Francisco, San Francisco, CA) Steven Cummings.
Clinical Centers: (Albert Einstein College of Medicine, Bronx, NY) Sylvia
Wassertheil-Smoller; (Baylor College of Medicine, Houston, TX) Jennifer Hays;
(Brigham and Women’s Hospital, Harvard Medical School, Boston, MA) JoAnn
Manson; (Brown University, Providence, RI) Annlouise R. Assaf; (Emory
University, Atlanta, GA) Lawrence Phillips; (Fred Hutchinson Cancer Research
Center, Seattle, WA) Shirley Beresford; (George Washington University Medical
Center, Washington, DC) Judith Hsia; (Harbor-UCLA Research and Education
Institute, Torrance, CA) Rowan Chlebowski; (Kaiser Permanente Center for
Health Research, Portland, OR) Evelyn Whitlock; (Kaiser Permanente Division
of Research, Oakland, CA) Bette Caan; (Medical College of Wisconsin,
Milwaukee, WI) Jane Morley Kotchen; (MedStar Research Institute/Howard
University, Washington, DC) Barbara V. Howard; (Northwestern University,
Chicago/Evanston, IL) Linda Van Horn; (Rush-Presbyterian St. Luke’s Medical
Center, Chicago, IL) Henry Black; (Stanford Prevention Research Center,
Stanford, CA) Marcia L. Stefanick; (State University of New York at Stony
Brook, Stony Brook, NY) Dorothy Lane; (The Ohio State University, Columbus,
OH) Rebecca Jackson; (University of Alabama at Birmingham, Birmingham,
AL) Cora E. Lewis; (University of Arizona, Tucson/Phoenix, AZ) Tamsen
Bassford; (University at Buffalo, Buffalo, NY) Jean Wactawski-Wende;
(University of California at Davis, Sacramento, CA) John Robbins; (University
of California at Irvine, Orange, CA) Allan Hubbell; (University of California at
Los Angeles, Los Angeles, CA) Howard Judd; (University of California at San
Diego, LaJolla/Chula Vista, CA) Robert D. Langer; (University of Cincinnati,
Cincinnati, OH) Margery Gass; (University of Florida, Gainesville/Jacksonville,
FL) Marian Limacher; (University of Hawaii, Honolulu, HI) David Curb;
(University of Iowa, Iowa City/Davenport, IA) Robert Wallace; (University of
Massachusetts/Fallon Clinic, Worcester, MA) Judith Ockene; (University of
Medicine and Dentistry of New Jersey, Newark, NJ) Norman Lasser; (University
of Miami, Miami, FL) Mary Jo O’Sullivan; (University of Minnesota,
Minneapolis, MN) Karen Margolis; (University of Nevada, Reno, NV) Robert
Brunner; (University of North Carolina, Chapel Hill, NC) Gerardo Heiss;
(University of Pittsburgh, Pittsburgh, PA) Lewis Kuller; (University of Tennessee,
Memphis, TN) Karen C. Johnson; (University of Texas Health Science Center,
San Antonio, TX) Robert Brzyski; (University of Wisconsin, Madison, WI)
Gloria E. Sarto; (Wake Forest University School of Medicine, Winston-Salem,
NC) Denise Bonds; (Wayne State University School of Medicine/Hutzel Hospital,
Detroit, MI) Susan Hendrix.
Funding to pay the Open Access publication charges for this article was
provided by the University of Pittsburgh Research and Development Funds.
Manuscript received October 14, 2005 ; revised March 8, 2006 ; accepted
March 28, 2006.
by guest on May 14, 2011