470 Articles | JNCI Vol. 103, Issue 6 | March 16, 2011
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Breast cancer is a heterogeneous disease that can be divided into
distinct subtypes based on patterns of gene expression (1–3) or
tumor marker staining (3–5). This biological heterogeneity trans-
lates to important clinical differences (2,4–9) and likely reflects
etiologic differences (10,11). One tumor subtype that has emerged
as being of particular clinical and public health significance is
triple-negative breast cancer. Triple-negative breast cancers, which
account for 10%–25% of invasive breast cancers (8,9,12–16), are
characterized by a lack of estrogen receptor (ER), progesterone
receptor (PR), and HER2 expression and typically exhibit a basal-
like pattern of gene expression (1,17,18). The triple-negative phe-
notype is associated with an aggressive pathology and poorer
prognosis than the predominant ER-positive (ER+) phenotype
(4,5,7–9), and there are currently no targeted therapies for the
treatment of triple-negative breast cancer. If the molecular profiles
of breast tumors are fixed at inception (11), distinct risk factors
would be expected to contribute towards triple-negative vs ER+
breast cancers. Nevertheless, even though many studies have
described risk factors for ER+ breast cancer (19–24), the etiology
and risk factor profile of triple-negative tumors remain poorly
Because triple-negative breast cancers are hormone receptor
negative, it is plausible that established risk factors for breast can-
cer overall that influence disease risk through hormonal
mechanisms could be differentially associated with risk of ER+ vs
triple-negative tumor subtypes. There are only a few studies (13–
16,25,26) that have assessed the role of potentially hormonally me-
diated risk factors for breast cancer overall in relation to risk of
triple-negative disease in particular. These studies have been
inconsistent and most have been limited by small numbers.
Reproductive History and Oral Contraceptive Use in Relation to
Risk of Triple-Negative Breast Cancer
Amanda I. Phipps, Rowan T. Chlebowski, Ross Prentice, Anne McTiernan, Jean Wactawski-Wende, Lewis H. Kuller,
Lucile L. Adams-Campbell, Dorothy Lane, Marcia L. Stefanick, Mara Vitolins, Geoffrey C. Kabat, Thomas E. Rohan, Christopher I. Li
Manuscript received June 28, 2010; revised January 13, 2011; accepted January 14, 2011.
Correspondence to: Amanda I. Phipps, PhD, Division of Public Health Sciences, Fred Hutchinson Cancer Research Center, 1100 Fairview Ave North,
M4-B402, PO Box 19024, Seattle, WA 98109-1024 (e-mail: email@example.com).
Background Triple-negative (ie, estrogen receptor [ER], progesterone receptor, and HER2 negative) breast cancer occurs
disproportionately among African American women compared with white women and is associated with a
worse prognosis than ER-positive (ER+) breast cancer. Hormonally mediated risk factors may be differentially
related to risk of triple-negative and ER+ breast cancers.
Methods Using data from 155 723 women enrolled in the Women’s Health Initiative, we assessed associations between
reproductive and menstrual history, breastfeeding, oral contraceptive use, and subtype-specific breast cancer
risk. We used Cox regression to evaluate associations with triple-negative (N = 307) and ER+ (N = 2610) breast
cancers and used partial likelihood methods to test for differences in subtype-specific hazard ratios (HRs).
Results Reproductive history was differentially associated with risk of triple-negative and ER+ breast cancers. Nulliparity
was associated with decreased risk of triple-negative breast cancer (HR = 0.61, 95% confidence interval [CI] =
0.37 to 0.97) but increased risk of ER+ breast cancer (HR = 1.35, 95% CI = 1.20 to 1.52). Age-adjusted absolute
rates of triple-negative breast cancer were 2.71 and 1.54 per 10 000 person-years in parous and nulliparous
women, respectively; by comparison, rates of ER+ breast cancer were 21.10 and 28.16 per 10 000 person-years
in the same two groups. Among parous women, the number of births was positively associated with risk of
triple-negative disease (HR for three births or more vs one birth = 1.46, 95% CI = 0.82 to 2.63) and inversely
associated with risk of ER+ disease (HR = 0.88, 95% CI = 0.74 to 1.04). Ages at menarche and menopause were
modestly associated with risk of ER+ but not triple-negative breast cancer; breastfeeding and oral contraceptive
use were not associated with either subtype.
Conclusion The association between parity and breast cancer risk differs appreciably for ER+ and triple-negative breast
cancers. These findings require further confirmation because the biological mechanisms underlying these dif-
ferences are uncertain.
J Natl Cancer Inst 2011;103:470–477
JNCI | Articles 471
However, multiple studies have reported an inverse association
between the number of months that a woman has spent breastfeed-
ing and her risk of triple-negative breast cancer (13,15,25–27).
Also, an increased risk of triple-negative breast cancer has been
reported among parous women (relative to nulliparous women)
(13–15), although an inverse association with parity has been
established for ER+ disease. Two studies have reported an
increased risk of triple-negative breast cancer among women who
have used oral contraceptives (16,27).
Using data from the Women’s Health Initiative (WHI) study,
we investigated associations between menstrual and reproductive
history, breastfeeding, use of oral contraceptives, and triple-
negative breast cancer risk among postmenopausal women. We
also assessed associations between these factors and risk of ER+
breast cancer for comparison.
Materials and Methods
The WHI is a longitudinal study of postmenopausal women,
including multiple concurrent randomized clinical trials and an
observational study. Details of the WHI study design have been
published previously (28,29). Briefly, postmenopausal women aged
50–79 years were recruited from 40 clinical centers across the
United States between October 1, 1993, and December 31, 1998.
Women were excluded if they had a medical condition that was
associated with less than 3 years of predicted survival time or were
unlikely to remain in the same geographic area for at least 3 years.
Additional eligibility criteria were imposed for participation in the
clinical trials component of the WHI study, but women who did
not meet these additional criteria or who were not interested in the
clinical trials were given the option to enroll in the observational
study. At the time of enrollment, all women provided written
informed consent for participation and completed a baseline ques-
tionnaire. Women in the clinical trials also received a clinical
breast examination and mammogram at baseline; women with
baseline examinations suspicious for breast cancer or with a history
of breast cancer were excluded from the clinical trials. Institutional
review boards at all participating institutions approved the WHI
Baseline information on demographic factors, medical history,
family history of breast cancer, physical activity, height, weight,
and mammography history were collected via self-administered
questionnaires. These questionnaires were also used to collect
detailed information on reproductive history, including parity, age
at first birth, breastfeeding, and menstrual history. Information
regarding use of oral contraceptives, postmenopausal hormone
therapy (HT), and other medications was collected through struc-
tured in-person interviews.
Mammography information and medical history, including
diagnosis of breast cancer, were updated on an annual (observa-
tional study) or semiannual (clinical trials) basis via mailed or
telephone-administered questionnaires. Per study protocol,
women in the clinical trials received clinical breast examinations
and mammography annually in the case of HT trials or biennially
in the case of dietary trials; these procedures were not part of the
study protocol for women in the observational study, although all
CONTEXT AND CAVEATS
Because triple-negative breast cancers are hormone receptor–
negative, unlike estrogen receptor–positive (ER+) breast cancers,
it is possible that hormones and reproduction have a different
influence on risk of such cancers.
Data from 307 women with triple-negative breast cancer, 2610
women with ER+ breast cancer, and 150 529 control subjects were
collected from the Women’s Health Initiative trials and observa-
tional study. Cox proportional hazards models were used to deter-
mine whether reproductive and menstrual history and use of oral
contraceptives were associated with risk of triple-negative and/or
ER+ breast cancers.
Women without children had increased risk of ER+ breast cancer,
but decreased risk of triple-negative breast cancer. However,
among women who had given birth, those with more children had
higher risk of triple-negative and lower risk of ER+ disease.
Menstrual history was modestly associated with risk of ER+ breast
cancer, but breastfeeding and use of oral contraceptives were not
associated with either disease.
Childbirth appears to have opposite influences on risk of triple-
negative vs ER+ breast cancer.
The Women’s Health Initiative study population was entirely of
postmenopausal women, so findings may not apply to younger
women. The findings should be confirmed with a larger population
of women with triple-negative breast cancer, particularly because
the mechanisms behind these associations are not understood.
From the Editors
women (in both the observational study and clinical trials) were
asked to report at each follow-up visit whether they had had a
mammogram since their last visit. Breast cancer diagnoses reported
by participants were verified locally by WHI physician adjudica-
tors. Medical records and pathology reports for locally confirmed
breast cancers were sent to the WHI Clinical Coordinating Center
for central adjudication and coding of ER, PR, and HER2 status.
In total, 161 808 women enrolled in the WHI, 93 676 in the
observational study and 68 132 in the clinical trials. After excluding
5239 women who had a history of breast cancer or mastectomy at
baseline and 846 women who were without follow-up information
for breast cancer diagnoses, the present analyses included 155 723
women. Over the course of a median of 7.9 years of follow-up,
invasive breast cancers were identified in 5194 women. Information
on ER, PR, and HER2 status was available for 4677 (90%), 4600
(89%), and 3139 (60%) of the 5194 women, respectively. Because
of variability in HER2 testing practices over the study period and
across institutions, we excluded 1334 ER+ case subjects with
unknown HER2 status from the ER+ group to make it more com-
parable with the triple-negative group. Of the 3116 case subjects
with complete tumor marker data, 307 (10%) were triple negative
472 Articles | JNCI Vol. 103, Issue 6 | March 16, 2011
and 2610 (84%) were ER+. Remaining cancers were either ER2
PR+ (N = 45) or ER2 PR2 HER2+ (N = 154).
We used Cox regression to assess associations between menstrual
history (ages at menarche and menopause), reproductive history
(parity and age at first birth), lifetime duration of breastfeeding,
oral contraceptive use (lifetime duration of use and age at first use),
and subtype-specific breast cancer risk. Proportional hazards as-
sumptions for all models were verified by testing for a nonzero
slope of the scaled Schoenfeld residuals on ranked failure times and
on the log of analysis time. Age at menopause was defined as either
the age at which a participant experienced her last menstrual
period, received a bilateral oophorectomy, or initiated use of
menopausal HT, whichever came first.
Separate regression models were constructed for the two out-
comes of interest (ie, triple-negative and ER+ breast cancer). In all
models, the time axis was defined as the time since random assign-
ment in the case of clinical trials and time since study enrollment
in the case of the observational study. Women diagnosed with in
situ breast cancer or with an invasive breast cancer other than the
model-specific outcome were censored at the time of diagnosis.
We also compared hazard ratios (HRs) with 95% confidence inter-
vals (CIs) for women with triple-negative and ER+ breast cancers
by using competing risks partial likelihood methods (30). P values
for comparisons were based on two-sided tests. Analyses were
performed using STATA SE version 10.1 (StataCorp, College
Analyses were adjusted for age at study enrollment or random
assignment (in 5-year intervals) and study arm through stratification
of the baseline hazards. We also adjusted for the following baseline
characteristics associated with overall breast cancer risk: race (non-
Hispanic white, Hispanic, African American, or other), educational
level (high school or less, vocational or training school, some college
or associate’s degree, or college graduate), family history of breast
cancer in first-degree relatives (yes, no), body mass index (in quar-
tiles), HT use (never use, exclusive use of estrogen-only HT, ever
use of combined estrogen–progestin HT), smoking history (never,
ever), history of mammography within the 2 years before baseline
(yes, no), and mammography during follow-up (time-varying
covariate, yes vs no mammogram since last study visit). Exposures of
interest were mutually adjusted for each other, with the exception
that we adjusted for parity (1, 2, or 3 or more births), age at first
birth (<20, 20–29, or ≥30 years), and breastfeeding (never, 1–6,
7–12, or >12 months) only in analyses restricted to parous women.
Our study included 2610 women with ER+ breast cancer, 307
women with triple-negative breast cancer, and 150 529 control
subjects whose demographic and tumor characteristics are provided
(Table 1). The women with triple-negative breast cancer were
younger, more likely to have a family history of breast cancer, and
had a higher grade and larger tumor size than the women with ER+
breast cancer. Women with triple-negative cancer were also more
likely to be African American. Age-adjusted incidence rates of
triple-negative breast cancer were 2.44 and 4.57 cancers per 10 000
person-years among non-Hispanic white and African American
women, and rates for ER+ disease were 23.30 and 14.15 cancers per
10 000 person-years for the same two groups, respectively.
Aspects of menstrual and reproductive history emerged as
being differentially associated with risk of ER+ and triple-negative
breast cancers (Table 2). Age at menarche was modestly inversely
associated and age at menopause was modestly positively associ-
ated with risk of ER+ breast cancer; these associations were not
evident for triple-negative breast cancer. More considerable differ-
ences were noted in associations between reproductive history and
risk of ER+ vs triple-negative breast cancer. Compared with
parous women, nulliparous women had an increased risk of ER+
breast cancer (HR = 1.35, 95% CI = 1.20 to 1.52) but a decreased
risk of triple-negative breast cancer (HR = 0.61, 95% CI = 0.37 to
0.97; Pcompeting risks = .02). Age-adjusted incidence rates for ER+ breast
cancer were 28.16 and 21.10 cancers per 10 000 person-years in
nulliparous and parous women, respectively. In comparison, age-
adjusted incidence of triple-negative disease was 1.54 and 2.71
cancers per 10 000 person-years in the same two groups. History
of pregnancy losses was not associated with risk of either subtype
(results not shown). Among parous women, the number of births
was inversely associated with ER+ breast cancer (HR = 0.88, 95%
CI = 0.74 to 1.04) but positively associated with triple-negative
breast cancer risk (HR for three births or more vs one birth = 1.46,
95% CI = 0.82 to 2.63), and age at first birth was positively associ-
ated with risk of ER+ cancers but not triple-negative disease;
however, differences between subtype-specific associations were
not statistically significant. Lifetime duration of breastfeeding was
not associated with risk of either subtype.
Analyses of lifetime duration of oral contraceptive use indicated
no association with risk of ER+ or triple-negative breast cancer,
with the exception of a modestly reduced risk of ER+ disease
among women who had used oral contraceptives for at least 10
years (HR = 0.80, 95% CI = 0.68 to 0.94) (Table 3). There was no
evidence of a difference in subtype-specific HR estimates (Pcompeting
risks = .49). Results were similar when stratified into broad age cate-
gories (50–59 years vs 60–79 years, data not shown). We also found
no association between the age at which a woman initiated oral
contraceptive use and risk of either subtype; however, these
analyses were limited in power because the vast majority of women
using oral contraceptives initiated use at or after age 25.
The results from this analysis are consistent with prior studies in
suggesting that reproductive factors play a different role in relation
to risk of triple-negative breast cancer vs ER+ breast cancer (13–
15). Specifically, we found that nulliparity was associated with a
39% lower risk of triple-negative breast cancer but a 35% higher
risk of ER+ disease in postmenopausal women. Among parous
women, we found that having multiple children was associated
with greater risk of triple-negative breast cancer but with lesser
risk of ER+ breast cancer. Although associations with other risk
factors were comparable across subtypes, differences in associa-
tions with parity are consistent with existing literature and with
hypothesized etiologic distinctions between ER+ and triple-nega-
JNCI | Articles 473
In the largest comparative study of triple-negative and other
breast cancer subtypes to date, which included 335 triple-negative
patients, Ma et al. (27) reported a decreased risk of ER+ breast
cancer (odds ratio = 0.55) among women who had at least four
pregnancies compared with nulligravid women but found no asso-
ciation with risk of triple-negative breast cancer (odds ratio = 1.00).
Smaller studies that included 78–187 triple-negative patients have
also reported no association between parity or age at first birth and
triple-negative breast cancer risk (16,25,26), although others have
found an increased risk of triple-negative disease in multiparous
In contrast to prior studies, we found no association between
breastfeeding and risk of triple-negative breast cancer. Five pre-
vious studies, conducted in diverse settings, have noted a statisti-
cally significantly lower risk of triple-negative breast cancer in
parous women who have ever breastfed a child (15), or who breastfed
Table 1. Distribution of demographic and tumor characteristics among case subjects with breast cancer and control subjects*
Characteristic Control subjects, n (%)
Case subjects, n (%)
ER+ Triple negative
Regional or distant
Poorly differentiated or anaplastic
Tumor size, mm
Age at random assignment or enrollment, y
Race and/or ethnicity
Hispanic or Latina
Other or unknown
≤High school diploma or GED
Vocational or training school
Some college or associate’s degree
Breast cancer family history
Menopausal hormone therapy use
Exclusive use of unopposed estrogen
Ever use of combined estrogen–progestin
BMI at baseline, kg/m2
N/A 15 (5)
50 400 (34)
67 505 (45)
32 624 (22)
124 008 (86)
13 675 (10)
33 928 (23)
15 347 (10)
41 508 (28)
58 622 (39)
116 663 (82)
25 648 (18)
56 534 (38)
47 871 (32)
46 101 (31)
75 933 (51)
72 648 (49)
37 377 (25)
37 318 (25)
37 326 (25)
37 207 (25)
* GED = General Educational Development; BMI = body mass index; ER+ = estrogen receptor–positive with known HER2 status; triple negative = ER2, PR2, and
474 Articles | JNCI Vol. 103, Issue 6 | March 16, 2011
for a cumulative duration of at least 4 (13), 6 (25,27), or 12 months
(26) (P < .05). Most of these studies have included premenopausal
and postmenopausal women, without stratification by meno-
pausal status. Although the one study conducted only in postmen-
opausal women (25) indicated a 50% lower risk of triple-negative
breast cancer in parous women who breastfed for at least 6
months compared with parous women who never breastfed, an-
other study (27) that stratified analyses by attained age indicated
that breastfeeding was more strongly associated with triple-
negative breast cancer among women aged 35–44 years than among
women aged 45–64 years. This latter finding is consistent with
other studies that did not stratify by tumor marker expression (31)
because it suggests that associations between breastfeeding and
breast cancer risk are most pronounced in premenopausal women.
Thus, it is possible that we observed no association between
breastfeeding and risk of triple-negative or ER+ breast cancers
because of the older age and postmenopausal status of the study
The age range of this study population may also explain why we
found no association between use of oral contraceptives and risk of
triple-negative breast cancer, as was previously suggested (16,27).
Dolle et al. reported a 4.7-fold increased risk of triple-negative
breast cancer in women who were younger than 40 years and used
oral contraceptives for 6 years or more. In the same age group,
they reported a 6.4-fold increased risk of triple-negative breast
cancer among women who initiated use of oral contraceptives
before age 18 compared with those who had never used them;
however, use of oral contraceptives was not associated with risk in
Table 2. Relationship between menstrual and reproductive history and risk of estrogen receptor-positive (ER+) and triple-negative
n (%)HR (95% CI)n (%)HR (95% CI)
Age at menarche,† y
Age at menopause,† y
≥1 full-term pregnancy
Age at first birth,‡ y
Duration of breastfeeding,‡ mo
32 816 (21)
82 463 (56)
34 660 (23)
0.87 (0.79 to 0.97)
0.89 (0.79 to 1.00)
0.79 (0.63 to 0.98)
0.85 (0.76 to 0.96)
1.13 (1.00 to 1.27)
1.35 (1.20 to 1.52)
0.88 (0.65 to 1.19)
0.96 (0.67 to 1.39)
0.91 (0.52 to 1.57)
0.76 (0.54 to1.06)
1.02 (0.68 to 1.52)
0.61 (0.37 to 0.97)
34 500 (24)
84 509 (59)
16 920 (12)
17 509 (11)
132 064 (89)
13 151 (10)
37 355 (28)
81 558 (62)
0.97 (0.81 to 1.15)
0.88 (0.74 to 1.04)
1.03 (0.88 to 1.20)
1.36 (1.10 to 1.67)
1.00 (0.89 to 1.12)
0.97 (0.83 to 1.13)
0.98 (0.85 to 1.13)
1.71 (0.96 to 3.07)
1.46 (0.82 to 2.63)
1.26 (0.82 to 1.94)
1.05 (0.53 to 2.06)
0.92 (0.66 to 1.27)
0.67 (0.41 to 1.10)
0.81 (0.53 to 1.26)
19 365 (16)
87 961 (74)
10 861 (9)
54 797 (42)
38 351 (29)
16 414 (13)
20 928 (16)
* ER+ = estrogen receptor–positive with known HER2 status; triple negative = ER2, PR2, and HER22; HR = hazard ratio.
† Adjusted for age, study arm, race, education level, family history of breast cancer, body mass index, hormone therapy use, smoking history, history of mammog-
raphy (at baseline), mammography during follow-up, age at menarche, age at menopause, nulliparity, and oral contraceptive use.
‡ Adjusted for age, study arm, race, education level, family history of breast cancer, body mass index, hormone therapy use, smoking history, history of mammog-
raphy (at baseline), mammography during follow-up, age at menarche, age at menopause, oral contraceptive use, parity, age at first birth, and breastfeeding.
§ P values are from two-sided tests that compared adjusted hazard ratios for the two breast cancer subtypes by competing risks partial likelihood methods.
JNCI | Articles 475
the upper age range of that study population (41–45 years) (16).
Recently, Ma et al. (27) reported an increased risk of triple-
negative breast cancer associated with use of oral contraceptives
but only among women aged 45–64 years who first used oral con-
traceptives before age 18. Our study included women aged 50–79
years and included no women with triple-negative cancers who had
used oral contraceptives before age 18. Thus, although our find-
ings suggest no association between use of oral contraceptives and
triple-negative breast cancer, inferences based on these results
should be restricted to postmenopausal women who initiated use
of oral contraceptives at a relatively older age.
As in previous reports (7,8,13,32), we found race and/or eth-
nicity to be a major factor associated with triple-negative breast
cancer: the triple-negative subtype accounted for 22% of breast
cancers in African American women compared with 9% in women
of other races and/or ethnicities. This difference is consistent with
another analysis within the WHI cohort (33), which reported that
African American women with breast cancer are more likely to
have poorly differentiated hormone receptor-negative disease than
non-Hispanic white women. That analysis found that differences
in breast cancer incidence rates between African American and
non-Hispanic white women were not fully explained by differences
in the distribution of established breast cancer risk factors. Based
on the present analysis, one possibility is that differences in risk
factor distributions do not explain differences in incidence rates
between these groups because risk factors for the triple-negative
subtype (which disproportionately occurs in African American
women) differ from those for ER+ disease. Although small
numbers prevented us from assessing subtype-specific associations
by race and/or ethnicity, we found that the prevalence of late age
at first birth was lower among African Americans with breast can-
cer (7.5%) than among non-Hispanic whites with breast cancer
(11.8%). The prevalence of nulliparity, however, was slightly
higher among African Americans with breast cancer (14.7% vs
13.6% in non-Hispanic whites).
In addition to the differences between the women with ER+
and triple-negative breast cancers that were observed here, hetero-
geneity within each of these groups of case subjects is also plau-
sible. Previous studies have indicated clinical and, to a lesser
extent, epidemiological differences between women who have
triple-negative tumors that express basal markers (ie, basal-like
cancers) and those who have triple-negative tumors that do not
express basal markers (ie, normal-like cancers) (3–5,13,14).
Differences in risk factor associations for ER+ breast cancer
according to PR and/or HER2 status have been suggested (13–
15,19,20,26,27). Heterogeneity within ER+ and triple-negative
subtypes could influence subtype-specific associations and compar-
isons between subtypes. However, heterogeneity within these
groups of case subjects is assumed to be less pronounced than the
differences between subtypes.
There are limitations to consider in interpreting these results.
Approximately 40% of case subjects had unknown HER2 status
and, therefore, did not contribute to either case group. Case sub-
jects with missing HER2 data were similar to other case subjects
with regard to all exposures and covariates. In sensitivity analyses,
we used multiple imputation to explore the potential impact of
censoring these observations at diagnosis and found almost no
difference from the results presented here. Additionally, some
misclassification of case groups may have resulted from the use of
tumor marker data from multiple laboratories across WHI clinical
centers because testing practices can vary; however, testing results
were reviewed centrally to minimize differences in classification
across institutions. Misclassification of exposure status is also plau-
sible because most exposures considered here occurred many years
Table 3. Relationship between oral contraceptive use and risk of estrogen receptor–positive (ER+) and triple-negative breast cancer*
n (%) HR (95% CI)†n (%)HR (95% CI)†
Lifetime duration of use of oral contraceptives, y
Age at first use of oral contraceptives, y
87 861 (58)
34 628 (23)
14 209 (9)
13 780 (9)
0.97 (0.87 to 1.08)
1.04 (0.89 to 1.20)
0.80 (0.68 to 0.94)
0.94 (0.85 to 1.03)
0.97 (0.81 to 1.15)
1.08 (0.67 to 1.72)
0.84 (0.65 to 1.18)
1.30 (0.88 to 1.93)
1.11 (0.72 to 1.70)
0.98 (0.73 to 1.32)
1.12 (0.72 to 1.76)
0.62 (0.15 to 2.60)
87 861 (58)
45 783 (30)
15 148 (10)
* ER+ = estrogen receptor-positive with known HER2 status; triple-negative = ER2, PR2, and HER22; HR = hazard ratio.
† Adjusted for age, study arm, race, education level, family history of breast cancer, body mass index, hormone therapy use, smoking history, history of mammog-
raphy (at baseline), mammography during follow-up, age at menarche, age at menopause, and nulliparity.
‡ P values are from two-sided tests that compared adjusted hazard ratios for the two breast cancer subtypes by competing risks partial likelihood methods.
476 Articles | JNCI Vol. 103, Issue 6 | March 16, 2011
before study enrollment or random assignment; although errors in
recall are likely, the prospective design of the WHI makes differ-
ential recall by case status unlikely. Lastly, as previously men-
tioned, these results may not be generalizable to younger women
because the WHI was restricted to postmenopausal women.
There are several important strengths to this analysis, including
its large size, prospective design, and completeness of follow-up
and exposure information. To date, few studies have examined
risk factors for triple-negative breast cancer, and many of these
have been underpowered (N = 78 to N = 335 women with triple-
negative disease). This analysis thus contributes to a sparse litera-
ture and provides further support for the distinct epidemiology of
triple-negative breast cancers.
It has been hypothesized that risk of ER+ breast cancer is posi-
tively associated with a woman’s cumulative lifetime exposure to
endogenous ovarian hormones (34); thus, aspects of reproductive
and menstrual history could influence risk by affecting the number
of ovulatory cycles a woman experiences over her lifetime. If hor-
monal mechanisms predominate in the relationships between re-
productive factors and breast cancer risk, it is not surprising that
our results suggest that nulliparity or low parity, late age at first
birth, early menarche, and late menopause are associated with risk
of ER+ breast cancer. Because triple-negative breast cancers are
hormone receptor negative, it seems plausible that risk factors
operating through hormonal mechanisms would be less important
in the etiology of triple-negative than ER+ breast cancers.
Nevertheless, it remains unclear why nulliparity (or low parity)
would be associated with a decreased risk of triple-negative breast
cancer. It is also unclear why the difference between ER+ and tri-
ple-negative subtypes would be limited to associations with parity
and age at first birth and not extend to differences in associations
with other aspects of reproductive history, such as duration of
breastfeeding, which also influence a woman’s cumulative lifetime
exposure to endogenous estrogens.
Given the poor prognosis associated with triple-negative breast
cancer, it remains important to identify the factors that influence a
woman’s risk of developing this subtype of disease and to further
characterize if and how such factors differ from risk factors for the
more predominant ER+ breast cancer subtype that has a better
1. Perou CM, Sorlie T, Eisen MB, et al. Molecular portraits of human breast
tumours. Nature. 2000;406(6797):747–752.
2. Sorlie T, Tibshirani R, Parker J, et al. Repeated observation of breast
tumor subtypes in independent gene expression data sets. Proc Natl Acad
Sci U S A. 2003;100(14):8418–8423.
3. Nielsen TO, Hsu FD, Jensen K, et al. Immunohistochemical and clinical
characterization of the basal-like subtype of invasive breast carcinoma.
Clin Cancer Res. 2004;10(16):5367–5374.
4. Carey LA, Perou CM, Livasy CA, et al. Race, breast cancer subtypes, and
survival in the Carolina Breast Cancer Study. JAMA. 2006;295(21):
5. Kim MJ, Ro JY, Ahn SH, et al. Clinicopathologic significance of the
basal-like subtype of breast cancer: a comparison with hormone receptor
and Her2/neu-overexpressing phenotypes. Hum Pathol. 2006;37(9):
6. Carey L, Dees E, Sawyer L, et al. The triple negative paradox: primary
tumor chemosensitivity of breast cancer subtypes. Clin Cancer Res. 2007;
7. Lund MJ, Trivers KF, Porter PL, et al. Race and triple negative threats to
breast cancer survival: a population-based study in Atlanta, GA. Breast
Cancer Res Treat. 2009;113(2):357–370.
8. Bauer KR, Brown M, Cress RD, et al. Descriptive analysis of estrogen
receptor (ER)-negative, progesterone receptor (PR)-negative, and HER2-
negative invasive breast cancer, the so-called triple-negative phenotype: a
population-based study from the California cancer Registry. Cancer.
9. Onitilo AA, Engel JM, Greenlee RT, et al. Breast cancer subtypes based
on ER/PR and Her2 expression: comparison of clinicopathologic features
and survival. Clin Med Res. 2009;7(1-2):4–13.
10. Sorlie T, Wang Y, Xiao C, et al. Distinct molecular mechanisms under-
lying clinically relevant subtypes of breast cancer: gene expression analyses
across three different platforms. BMC Genomics. 2006;7:127.
11. Lacroix M, Toillon RA, Leclercq G. Stable ’portrait’ of breast tumors
during progression: data from biology, pathology and genetics. Endocr
Relat Cancer. 2004;11(3):497–522.
12. Swain SM. Triple-negative breast cancer: metastatic risk and role of plat-
inum agents. 44th Annual Meeting of the American Society of Clinical
Oncology. 2008; Chicago, IL.
13. Millikan RC, Newman B, Tse CK, et al. Epidemiology of basal-like breast
cancer. Breast Cancer Res Treat. 2008;109(1):123–139.
14. Yang XR, Sherman ME, Rimm DL, et al. Differences in risk factors for
breast cancer molecular subtypes in a population-based study. Cancer
Epidemiol Biomarkers Prev. 2007;16(3):439–443.
15. Xing P, Li J, Jin F. A case-control study of reproductive factors associated
with subtypes of breast cancer in Northeast China. Med Oncol.
16. Dolle JM, Daling JR, White E, et al. Risk factors for triple-negative breast
cancer in women under the age of 45 years. Cancer Epidemiol Biomarkers
17. Bertucci F, Finetti P, Cervera N, et al. How basal are triple-negative
breast cancers? Int J Cancer. 2008;123(1):236–240.
18. Kreike B, van Kouwenhove M, Horlings H, et al. Gene expression pro-
filing and histopathological characterization of triple-negative/basal-like
breast carcinomas. Breast Cancer Res. 2007;9(5):R65.
19. Potter JD, Cerhan JR, Sellers TA, et al. Progesterone and estrogen recep-
tors and mammary neoplasia in the Iowa Women’s Health Study: how
many kinds of breast cancer are there? Cancer Epidemiol Biomarkers Prev.
20. Ma H, Bernstein L, Pike MC, Ursin G. Reproductive factors and breast
cancer risk according to joint estrogen and progesterone receptor
status: a meta-analysis of epidemiological studies. Breast Cancer Res. 2006;
21. Huang W, Newman B, Millikan R, Schell M, Hulka B, Moorman P.
Hormone-related factors and risk of breast cancer in relation to estrogen
receptor and progesterone receptor status. Am J Epidemiol. 2000;151(7):
22. Rosenberg L, Einarsdottir K, Friman E, et al. Risk factors for hormone
receptor-defined breast cancer in postmenopausal women. Cancer
Epidemiol Biomarkers Prev. 2006;15(12):2482–2488.
23. Ursin G, Bernstein L, Lord SJ, et al. Reproductive factors and subtypes of
breast cancer defined by hormone receptor and histology. Br J Cancer.
24. Althuis MD, Fergenbaum JH, Garcia-Closas M, Brinton LA, Madigan
MP, Sherman ME. Etiology of hormone receptor-defined breast cancer: a
systematic review of the literature. Cancer Epidemiol Biomarkers Prev.
25. Phipps AI, Malone KE, Porter PL, et al. Reproductive and hormonal risk
factors for postmenopausal luminal, HER-2-overexpressing, and triple-
negative breast cancer. Cancer. 2008;113(7):1521–1526.
26. Trivers KF, Lund MJ, Porter PL, et al. The epidemiology of triple-
negative breast cancer, including race. Cancer Causes Control. 2009;20(7):
27. Ma H, Wang Y, Sullivan-Halley J, et al. Use of four biomarkers
to evaluate the risk of breast cancer subtypes in the women’s contra-
ceptive and reproductive experiences study. Cancer Res. 2010;70(2):
jnci.oxfordjournals.org Download full-text
JNCI | Articles 477
28. The Women’s Health Initiative Study Group. Design of the Women’s
Health Initiative clinical trial and observational study. Control Clin Trials.
29. Hays J, Hunt JR, Hubbell FA, et al. The Women’s Health Initiative re-
cruitment methods and results. Ann Epidemiol. 2003;13(9)(suppl):S18–S77.
30. Kalbfleisch JD, Prentice RL. The Statistical Analysis of Failure Time Data.
2nd ed. New York, NY: John Wiley & Sons, Inc; 2002.
31. Lipworth L, Bailey LR, Trichopoulos D. History of breast-feeding in
relation to breast cancer risk: a review of the epidemiologic literature.
J Natl Cancer Inst. 2000;92(4):302–312.
32. Stark A, Kapke A, Schultz D, et al. Advanced stages and poorly differenti-
ated grade are associated with an increased risk of HER2/neu positive
breast carcinoma only in white women: findings from a prospective cohort
study of African-American and white-American women. Breast Cancer Res
33. Chlebowski RT, Chen Z, Anderson GL, et al. Ethnicity and breast cancer:
factors influencing differences in incidence and outcome. J Natl Cancer
34. Clavel-Chapelon F; E3N-EPIC Group. Differential effects of reproduc-
tive factors on the risk of pre- and postmenopausal breast cancer: results
from a large cohort of French women. Br J Cancer. 2002;86(5):723–727.
National Heart, Lung, and Blood Institute, National Institutes of Health, U.S.
Department of Health and Human Services (contracts N01WH22110, 24152,
32100-2, 32105-6, 32108-9, 32111-13, 32115, 32118-32119, 32122, 42107-
26, 42129-32, and 44221). National Cancer Institute (T32 CA09168 and
R25-CA94880 to A.I.P).
The National Institutes of Health had no role in the design and conduct of
the study; in the collection, analysis, and interpretation of the data; or in the
preparation, review, approval, or submission of the manuscript. The contents
of this publication are solely the responsibility of the authors and do not nec-
essarily represent the official views of the National Cancer Institute or the
National Institutes of Health. R. T. Chlebowski is a consultant for Novartis,
Pfizer, Astra-Zeneca, and Amgen and has received honoraria from Novartis and
Affiliations of authors: Division of Public Health Sciences, Fred Hutchinson
Cancer Research Center, Seattle, WA (AIP, RP, AM, CIL); Department of
Medicine, Los Angeles Biomedical Research Institute at Harbor-UCLA
Medical Center, Los Angeles, CA (RTC); Department of Social and
Preventive Medicine, School of Public Health and Health Professions,
University at Buffalo, Buffalo, NY (JW-W); Department of Epidemiology,
Graduate School of Public Health, University of Pittsburgh, Pittsburgh, PA
(LHK); Department of Oncology, Lombardi Comprehensive Cancer Center,
Georgetown University, Washington, DC (LLA-C); Department of Preventive
Medicine, State University of New York at Stony Brook, Stony Brook, NY
(DL); Stanford Prevention Research Center, Department of Medicine,
School of Medicine, Stanford University, Palo Alto, CA (MLS); Department
of Epidemiology and Prevention, Division of Public Health Sciences, Wake
Forest University, Winston-Salem, NC (MV); Department of Epidemiology
and Population Health, Albert Einstein College of Medicine, Bronx, NY (GCK,