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Advance Access publication April 29, 2008 Conjugated Equine Estrogens and Breast Cancer Risk in the Women’s HealthInitiative Clinical Trial and Observational Study Ross L. Prentice1, Rowan T. Chlebowski2, Marcia L. Stefanick3, JoAnn E. Manson4, Robert D.
Langer5, Mary Pettinger1, Susan L. Hendrix6, F. Allan Hubbell7, Charles Kooperberg1, Lewis H.
Kuller8, Dorothy S. Lane9, Anne McTiernan1, Mary Jo O’Sullivan10, Jacques E. Rossouw11, andGarnet L. Anderson1 1 Division of Public Health Sciences, Fred Hutchinson Cancer Research Center, Seattle, WA.
2 Los Angeles Biomedical Research Institute at Harbor-UCLA Medical Center, Los Angeles, CA.
3 Stanford Prevention Research Center, School of Medicine, Stanford University, Stanford, CA.
4 Division of Preventive Medicine, Brigham and Women’s Hospital, Harvard Medical School, Boston, MA.
5 Outcomes Research Institute, Geisinger Health System, Danville, PA.
6 Department of Obstetrics and Gynecology, Wayne State University, Detroit, MI.
7 Department of Medicine, University of California, Irvine, CA.
8 Department of Epidemiology, Graduate School of Public Health, University of Pittsburgh, Pittsburgh, PA.
9 Department of Preventive Medicine, State University of New York, Stony Brook, NY.
10 Department of Obstetrics and Gynecology, University of Miami, Miami, FL.
11 National Heart, Lung, and Blood Institute, Bethesda, MD.
Received for publication February 23, 2007; accepted for publication October 2, 2007.
The Women’s Health Initiative randomized controlled trial found a trend (p ¼ 0.09) toward a lower breast cancer risk among women assigned to daily 0.625-mg conjugated equine estrogens (CEEs) compared with placebo, incontrast to an observational literature that mostly reports a moderate increase in risk with estrogen-alone prepa-rations. In 1993–2004 at 40 US clinical centers, breast cancer hazard ratio estimates for this CEE regimen werecompared between the Women’s Health Initiative clinical trial and observational study toward understanding thisapparent discrepancy and refining hazard ratio estimates. After control for prior use of postmenopausal hormonetherapy and for confounding factors, CEE hazard ratio estimates were higher from the observational study comparedwith the clinical trial by 43% (p ¼ 0.12). However, after additional control for time from menopause to first use ofpostmenopausal hormone therapy, the hazard ratios agreed closely between the two cohorts (p ¼ 0.82). For womenwho begin use soon after menopause, combined analyses of clinical trial and observational study data do not provideclear evidence of either an overall reduction or an increase in breast cancer risk with CEEs, although hazard ratiosappeared to be relatively higher among women having certain breast cancer risk factors or a low body mass index.
breast neoplasms; clinical trial; cohort studies; estrogens; hormone replacement therapy; postmenopause Abbreviations: CEE, conjugated equine estrogen; CI, confidence interval; HT, postmenopausal hormone therapy;WHI, Women’s Health Initiative.
Editor’s note: An invited commentary on this article is The Women’s Health Initiative (WHI) randomized con- trolled trial of daily use of 0.625 mg of conjugated equine Correspondence to Dr. Ross L. Prentice, Division of Public Health Sciences, Fred Hutchinson Cancer Research Center, 1100 Fairview AvenueNorth, P.O. Box 19024, Seattle, WA 98109-1024 (e-mail: rprentic@fhcrc.org).
estrogens (CEEs) by 10,739 posthysterectomy US women tended to provide risk factor information on major causes of stopped early in 2004 primarily because of an elevation in morbidity and mortality and to serve as a secular control for stroke risk (1, 2). The trial yielded a nonsignificantly lower the clinical trials. All women provided written informed con- incidence of invasive breast cancer in the active hormone sent for their respective WHI activities and supplied a baseline group (3), with a hazard ratio of 0.80 (95 percent confidence fasting blood specimen, a medications and dietary supple- interval (CI): 0.62, 1.04; p ¼ 0.09) over an average 7.1-year ments inventory, and common core questionnaires (7, 9).
follow-up period. This hazard ratio estimate compares with Information on lifetime hormone use was obtained from generally higher hazard ratios from an extensive observa- clinical trial and observational study women at baseline by a trained interviewer, assisted by a structured questionnaire For example, the UK Million Women Study (5) reported a and chart displaying color photographs of various hormone hazard ratio of 1.30 (95 percent CI: 1.21, 1.40) for estrogen- preparations. For HT, detailed information was obtained on alone regimens, with little evidence of hazard ratio variation the preparation, estrogen and progestin doses, schedule, and among regimens involving differing estrogens. In the WHI route of administration. The age at starting and stopping CEE trial, much of the evidence for a possibly reduced each preparation was recorded. Estrogen-alone use was de- breast cancer hazard ratio (3) arose from women who had fined as use of prescription oral or transdermal preparations not previously used postmenopausal hormone therapy (HT), for at least 3 months, whereas estrogen plus progestin use many of whom were years past menopause at the time of was defined similarly for estrogen plus oral progestin, in- trial enrollment. The hazard ratio was 0.65 (95 percent CI: cluding preparations used continuously or intermittently.
0.46, 0.92) among women without prior HT compared with Women using HT at baseline were required to undergo 1.02 (95 percent CI: 0.70, 1.50) among women with prior a 3-month washout period prior to randomization in the HT (p ¼ 0.09 for difference). In spite of only 237 incident CEE trial. Women without a uterus were potentially eligible cases, this trial was able to identify higher hazard ratios for this trial of daily use of 0.625 mg of CEE or matched among subsets of women at comparatively high risk, includ- placebo. There were no restrictions on hormone therapy use ing those with an elevated 5-year Gail model (6) risk score for observational study participants.
(p ¼ 0.01), those having one or more first-degree relatives This article is based on data from a clinical trial and an with breast cancer (p ¼ 0.01), and those having a personal observational study subcohort of women enrolled at 40 US history of benign breast disease (p ¼ 0.005).
clinical centers. The ‘‘gap time’’ from menopause to first Here, we compare results from the CEE trial with corre- use of HT emerged in preliminary analyses as a useful factor sponding results from the WHI observational study with for explaining hazard ratio patterns, so the clinical trial a goal of identifying reasons for any hazard ratio discrep- analyses presented here excluded women of an unknown ancy. If results from the two cohorts are in good agreement age at menopause or having unknown prior HT information.
following provision for differences in the characteristics and Following this exclusion, 4,493 (84.6 percent) of the women HT exposure patterns of participating women, then analysis assigned to CEE and 4,596 (84.7 percent) of the women of combined data from the two sources may help to clarify assigned to placebo remained in the clinical trial subcohort the breast cancer effects of CEE, especially among recently The corresponding observational study subcohort com- The WHI observational study includes 93,676 postmen- prised 17,437 women who were either using the same daily opausal women enrolled from the same populations as the 0.625-mg CEE regimen (9,336 women) or were not using WHI clinical trial (7) over essentially the same time period any HT (8,101 women) at the time of enrollment in the (1993–1998). Many elements of the protocol were common observational study. To enhance comparability with the clin- to the two WHI components, including baseline question- ical trial, observational study subcohort women were re- naire and interview data collection and the major elements quired to be posthysterectomy, to have no personal history of breast cancer, and to have had a mammogram within2 years prior to enrollment. As with the clinical trial sub-cohort, women were also required to be of a known age atmenopause and to have prior HT data. Finally, women in this observational study subcohort were required to have known values for a list of potential confounding factors de-scribed below (figure 1).
Detailed WHI recruitment methods and eligibility criteria have been published previously (8). Eligible women were 50–79 years of age at screening, were postmenopausal, hadno medical condition associated with a predicted survival of Clinical outcomes were reported semiannually in the clin- less than 3 years, and were likely to be residing in the same ical trial and annually in the observational study. Initial geographic area for at least 3 years. Additional CEE trial reports of outcomes were ascertained by self-administered exclusion criteria involved safety, adherence, and retention questionnaire. Breast cancer occurrences were confirmed concerns and included a personal history of invasive or non- by review of medical records and pathology reports by invasive breast cancer. Women ineligible for, or not interested physician-adjudicators at the local clinical centers. All in, the WHI clinical trial were given the opportunity to enroll cases were subsequently classified (10) at the Clinical Co- in the observational study. The observational study was in- ordinating Center by using the National Cancer Institute’s Primary data analyses used time-to-event methods based on the Cox regression procedure (11), with time from ran-domization in the clinical trial and time from enrollment inthe observational study as the basic time variable. Incidence rates of invasive breast cancer during follow-up were strat- ified on baseline age in 5-year categories and on clinical trialor observational study cohort.
Disease events in the CEE trial were included through February 29, 2004, when women stopped taking study pills, giving an average 7.1 years of follow-up. Follow-up time in the observational study subcohort was included throughDecember 15, 2004, to give an equivalent average follow-uptime of 7.1 years.
Confounding in the observational study was addressed by including breast cancer risk factors, collected at baseline, in the Cox regression model. Because such factors are inde-pendent of treatment assignment, they were not included in Potential confounding factors in observational study anal- yses (in addition to stratification on baseline age) included age(linear), body mass index (<25, 25–29, 30–34, >34 kg/m2,plus linear), education (high school or less, beyond high school, college degree), smoking (never, past, current), alco- hol consumption (never, past, <1/week, 1–7/week, >7/week),general health (fair/poor, good/very good/excellent), physicalactivity in metabolic equivalent units/week (0–3.75, 3.76– Numbers of US women in the Women’s Health Initiative 8.75, 8.76–17.5, >17.5), family history of breast cancer observational study meeting selection criteria, United States, 1993–2004. CEE, conjugated equine estrogen; HT, postmenopausal (yes, no), 5-year Gail model (6) breast cancer risk percent- age (<1.25, 1.25–1.74, >1.74, plus linear), and bilateraloophorectomy (yes, no). This rather extensive list aimedto control confounding as thoroughly as practical, withoutintroducing sparse-data biases (12).
Cox model hazard ratio estimates were calculated sepa- Surveillance, Epidemiology, and End Results coding system rately for less than 2, 2–5, and more than 5 years from CEE initiation, with proportional hazards within these time peri- Yearly mammography and clinical breast examination ods. Time from CEE initiation was defined as time from were required in the CEE trial, and study medications were randomization for women randomized to CEE in the clinical withheld if these procedures were not completed. Mammo- trial. Women who had not used any HT before randomiza- gram reports were obtained from performance sites and tion were classified as ‘‘no prior HT,’’ while all other clinical were reviewed locally and coded for recommendation.
trial women were included in a ‘‘prior HT’’ group. Time Mammograms with suspicious abnormalities or highly sug- from CEE initiation among CEE users in the observational gestive of malignancy required clearance before additional study was defined as the sum of the duration of the ongoing study medication was dispensed. In the observational study, daily 0.625-mg CEE episode at enrollment plus time since annual data collection updated each woman’s mammogram enrollment. A usage gap of 1 year or longer was required to history, and the WHI did not intervene regarding the mam- define a new CEE episode. CEE users who had used any HT mography practices of participating women.
prior to the beginning of the CEE episode ongoing at obser- Information on the use of HT was updated semiannually vational study enrollment were classified as prior HT.
in the clinical trial and annually in the observational study.
Women in the nonuser group in the observational study wereclassified as prior HT if they had used any HT prior to Disease incidence rates in the Cox model were also strat- Age at menopause was defined as the age at which ified on prior HT, and confounding factor coefficients (ob- a woman last had menstrual bleeding, had a bilateral oopho- servational study only) were estimated separately for the rectomy, or began using HT. Any such age greater than prior HT and no prior HT groups. Additional potential con- 60 years was recoded as 60 years. Age at first use of HT founding factors were included for the prior HT stratum as was defined both for women who had used any HT prior to follows: prior estrogen-alone use in years (none vs. each of WHI enrollment and for women whose first use of HT was <5, 5–10, >10) and prior estrogen plus progestin use in the active treatment in the clinical trial. The gap time from years (none vs. each of <5, 5–10, >10).
menopause to (first) HT use was the difference between A product term between a CEE and an observational study (vs. clinical trial) indicator variable in the log-hazard ratio Invasive breast cancer incidence rates in the US Women’s Health Initiative CEE* clinical trial and corresponding observational study subcohort according to prioruse of hormone therapy, 1993–2004 * CEE, 0.625 mg/day of conjugated equine estrogens; HT, postmenopausal hormone therapy.
y Prior HT was defined relative to the baseline CEE episode for CEE users in the observational study and relative to Women’s Health Initiative enrollment for other women.
z The observational study subcohort comprises women with a hysterectomy, without a prior breast cancer diagnosis, with a mammogram in the 2 years prior to observational studyenrollment, and either using the daily 0.625-mg CEE regimen studied in the clinical trial or notusing any HT at the time of enrollment.
§ Only invasive breast cancer diagnoses that occurred within 2 years of the most recent { Age adjusted to the 5-year age distribution in the clinical trial cohort.
enabled hazard ratios in the observational study to differ by Confounding in the observational study could have con- a multiplicative factor from those in the clinical trial. Estima- tributed to these patterns. In addition, hazard ratio compar- tion of this ‘‘CEE in the observational study/CEE in the isons between the clinical trial and observational study need clinical trial’’ hazard ratio factor provides an overall test of to acknowledge the different durations of CEE use in the two agreement between CEE effects in the two cohorts, and the cohorts, since most CEE users were some years into their inclusion of this factor in data analysis provides for residual baseline episode of CEE at the time of enrollment in the confounding in the observational study.
observational study. Table 2 includes the numbers of obser- In both the clinical trial and observational study cohorts, vational study women who developed breast cancer during hazard ratios were standardized for mammographic screen- follow-up in the time periods less than 2, 2–5, and more than ing patterns during WHI follow-up by censoring the follow- 5 years from CEE initiation, along with corresponding num- up for a woman when she first exceeded 2 years without bers from the clinical trial. Much of the information from the observational study for assessing CEE effects pertains to the In some analyses, hazard ratios among women who were more than 5 years from initiation category, where the clinical adherent to CEE were estimated by censoring the follow-up trial information is comparatively limited, but the overlap in period for a woman 6 months after she stopped taking CEE time from CEE initiation distributions between the two co- if in a user group or 6 months after initiating any HT if in horts was sufficient to allow a meaningful comparison be- a nonuser group. The 6-month period was included to avoid tween corresponding hazard ratio estimates.
HT changes related to diagnostic work-up from inappropri- The hazard ratio estimates (and 95 percent confidence intervals) in table 2 arose from Cox model (11) analysis In this paper, nominal 95 percent confidence intervals are of combined clinical trial and observational study data that presented for hazard ratio parameters. In addition, two-sided included confounding factors in the observational study (re- significance tests (p values) are presented.
fer to the Materials and Methods section). These analysesalso included a hazard ratio interaction between CEE and cohort that led to an overall ratio of the CEE hazard ratio inthe observational study to the CEE hazard ratio in the clin- Table 1 shows the numbers of women, their mean ages, ical trial estimated at 1.43 (95 percent CI: 0.91, 2.26). This and the number of incident breast cancers in the clinical trial 43 percent larger hazard ratio estimate in the observational and observational study subcohorts, separately by prior HT study compared with that in the clinical trial (p ¼ 0.12) status. The age-adjusted incidence rate ratios for CEE users suggests that confounding factors and different distributions compared with nonusers were similar from the clinical trial of time from CEE initiation may not fully explain differen- and observational study among women having prior HT but tial hazard ratios for CEE use between the two cohorts.
were lower in the clinical trial than in the observational Women without prior HT who enrolled in the CEE trial were often many years past menopause at the time of Breast cancer hazard ratio estimates for CEE* according to prior postmenopausal hormone therapy status and years from hormone therapy initiation forUS women, 1993–2004y * CEE, 0.625 mg/day of conjugated equine estrogens; HT, postmenopausal hormone therapy; HR, hazard ratio; CI, confidence interval.
y CEE in the observational study/CEE in the clinical trial: HR ¼ 1.43, 95% CI: 0.91, 2.26.
z Prior HT was defined relative to the baseline episode for CEE users in the observational study and relative to Women’s Health Initiative enrollment otherwise. Confounding factors in theobservational study were controlled separately in the prior HT and no prior HT groups and arelisted in the Materials and Methods section of the text.
§ No. of cases among CEE users in the clinical trial/no. of cases among CEE users in the observational study that contribute to the hazard ratio estimate.
randomization, whereas many CEE users in the observa- rate hazard ratios (for CEE use) according to prior HT and tional study were comparatively few years beyond meno- gap time from menopause to first use of HT (<5 vs. 5 years).
pause at the beginning of their baseline HT episode.
These analyses (table 4) provided a suggestion (p ¼ 0.20) of Similarly, women with prior HT in either the clinical trial lower hazard ratios among women having longer (5 years) or the observational study mostly initiated HT within a few gap times as a possible explanation for lower hazard ratios years following menopause. Table 3 shows the stark contrast between clinical trial women without prior HT and the other Gap time was next considered as a factor to explain three groups regarding the distribution of gap time from apparent differences between hazard ratios in the observa- menopause to first use of HT in the CEE user groups. Note, tional study and in the clinical trial. To do so, a product term for example, that there were only four breast cancer cases was included on the Cox model log-hazard ratio between among women without prior HT who were randomized to a CEE indicator and gap time. To avoid an undue influence CEE within 5 years following menopause.
by some very long gap times, gap times of more than 15 To examine the effect of gap time distribution on clinical years were recoded as 15 years. Table 5 shows estimated trial results, the clinical trial data were analyzed with sepa- hazard ratios and 95 percent confidence intervals for women Distribution of gap time from menopause to first use of postmenopausal hormones among CEE* users in the clinical trial and observational study according toprior use of postmenopausal hormone therapy by US women, 1993–2004 Gap time (years) from menopause to first use of HT* * CEE, 0.625 mg/day of conjugated equine estrogens; HT, postmenopausal hormone therapy.
y Prior HT was defined relative to the ongoing CEE episode at Women’s Health Initiative enrollment in the observational study and relative to randomization to CEE in the clinical trial.
z Women were selected to have a known time from menopause to first use of HT (refer to the Materials and Methods section of the text).
Invasive breast cancer hazard ratios for CEE* by Breast cancer hazard ratio estimates for CEE* years from menopause to first hormone therapy use in according to prior postmenopausal hormone therapy and years the Women’s Health Initiative clinical trial, United States, from CEE initiation among US women who initiated CEE at No. of years from menopause to first HT* use * CEE, 0.625 mg/day of conjugated equine estrogens; HT, post- * CEE, 0.625 mg/day of conjugated equine estrogens; HT, post- menopausal hormone therapy; HR, hazard ratio; CI, confidence menopausal hormone therapy; HR, hazard ratio; CI, confidence y Hazard ratios (and 95% confidence intervals) from Cox model y CEE in the observational study/CEE in the clinical trial: HR ¼ analyses that stratified on baseline age (5-year categories). Numbers of women and breast cancer cases contributing to each hazard ratio z Prior HT was defined relative to the baseline episode for CEE users in the observational study and relative to Women’s Health z Prior HT was defined relative to enrollment in the clinical trial.
Initiative enrollment otherwise. Confounding factors in the observa-tional study were controlled separately in the prior HT and no prior HTgroups and are listed in the Materials and Methods section of the text.
Refer to table 2 for numbers of breast cancer cases in the clinical trial who began CEE use immediately following menopause (gap and observational study in this table. Corresponding hazard ratioestimates for women who first initiate CEE following x years after time of 0). The dependence of the CEE hazard ratio on this menopause (up to 15) can be obtained by multiplying those in the gap time variable was significant (p ¼ 0.03) in these anal- yses. Corresponding hazard ratio estimates for women whoinitiated CEE, for example, 5 years following menopause,under this statistical model, would have been lower thanthose shown in table 5 by a factor of 0.85 (95 percent CI: each cell defined by gap years from menopause to CEE 0.73, 0.98). The ratio of the hazard ratio for CEE use from initiation and years from CEE initiation. The data are quite the observational study to that from the clinical trial was sparse in some cells, and confidence intervals may be in- 1.07 (95 percent CI: 0.60, 1.93; p ¼ 0.82), indicating excel- accurate. Nevertheless, a pattern of lower hazard ratios lent agreement overall between the clinical trial and obser- among women whose gap time was greater than 5 years is vational study after this accommodation of gap time.
evident. Hazard ratios among women whose gap times were Further analyses applied this same hazard ratio model sep- less than 5 years did not suggest a breast cancer risk re- arately in the clinical trial and observational study, and no duction with CEE. As shown in the lower part of table 6, evidence was found for a difference between cohorts in these patterns persisted among women adherent to their either gap time coefficients (p ¼ 0.56) or overall hazard CEE user or nonuser classification (refer to the Materials ratio functions (likelihood ratio p ¼ 0.92). The hazard ratio for CEE use also decreased with increasing gap time (p ¼ Given the good agreement between clinical trial and ob- 0.03) in a corresponding analysis of observational study data servational study hazard ratios shown in table 5, it was of interest to use the combined clinical trial and observational Some additional analyses were carried out to elucidate study data to reexamine the previously mentioned interac- the interpretation of the gap time association with hazard tions (3) of the CEE hazard ratio with other factors. Adding ratio, as follows: An interaction of CEE with a linear term in interaction factors one at a time to the table 5 analysis re- years from CEE initiation was added to the analysis to allow sulted in estimated hazard ratios that increased by a factor of for any residual duration effects beyond the categories given 1.14 (95 percent CI: 1.01, 1.29; p ¼ 0.04) with a one-unit in table 5 and was found to not be significant (p ¼ 0.65; increase in 5-year Gail model (6) breast cancer risk; in- hazard ratio ¼ 1.00 for this factor, 95 percent CI: 0.98, creased by a factor of 1.42 (95 percent CI: 1.00, 2.02; p ¼ 1.01). Similarly, age at WHI enrollment was not significant 0.05) among women with a history of benign breast disease; as a potential additional interaction factor (p ¼ 0.84; hazard increased by a factor of 1.27 (95 percent CI: 0.87, 1.84; p ¼ ratio ¼ 1.00, 95 percent CI: 0.98, 1.02), nor was age at HT 0.21) among women with a first-degree relative with breast initiation (p ¼ 0.33; hazard ratio ¼ 1.01, 95 percent CI: cancer; and decreased by a factor of 0.97 (95 percent CI: 0.99, 1.03). In each of these analyses, time from menopause 0.95, 1.00; p ¼ 0.03) for a one-unit increase in body mass to HT initiation remained significantly associated with the CEE hazard ratio (p < 0.005) in the presence of the otherfactor, pointing to the value of gap time as a relevant timescale to characterize CEE effects on breast cancer risk.
The upper part of table 6 presents a more empirical, less model-dependent view of CEE hazard ratios among women Preliminary analysis of WHI observational study data on without prior HT, with a separate hazard ratio estimate in postmenopausal estrogen-alone regimens in relation to Breast cancer hazard ratios according to gap years from menopause to CEE* use, and years from CEE initiation among US women without prior postmenopausal hormone therapy, from combinedanalysis of clinical trial and observational study data, 1993–2004y Gap time (years) from menopause to first use of CEE * CEE, conjugated equine estrogen; HR, hazard ratio; CI, confidence interval.
y No prior HT was defined relative to the beginning of the baseline CEE episode among CEE users in the observational study and relative to Women’s Health Initiative enrollment otherwise. Confounding factors in theobservational study are listed in the Materials and Methods section of the text.
z No. of breast cancer cases in the clinical trial/no. of breast cancer cases in the observational study that contribute to the hazard ratio estimate.
invasive breast cancer yielded a hazard ratio estimate of 1.28 trial and observational study data were combined, the hazard for estrogen-alone users versus nonusers of HT after control ratios reported in this article were not precisely determined.
for the set of confounding factors used in this presentation However, the analysis presented here suggests agreement (refer to the Materials and Methods section). This hazard between hazard ratios from the clinical trial and observa- ratio estimate was 79 percent larger (p < 0.01) than the tional study after control for gap time, and they give results corresponding estimate (HR ¼ 0.71) from the CEE trial.
generally consistent with an extensive related observational As shown in table 2, this discrepancy was reduced to 43 literature (4, 5). Hence, observational studies would seem to percent (p ¼ 0.12) after control for mammographic screen- be a reasonable source for more precise estimates of CEE ing patterns prior to and following WHI enrollment, restrict- effects. The fact that hazard ratios depended on gap time, as ing the estrogen-alone user group in the observational study well as mammographic screening pattern and other factors to women using the same daily 0.625-mg CEE regimen (e.g., body mass index) in analysis of WHI data, suggests studied in the clinical trial and controlling for time from that these factors should be considered in observational CEE initiation. These factors, particularly mammographic screening patterns, should be carefully controlled in obser- In a separate article (13), we present corresponding anal- vational studies of hormone therapy effects on breast cancer.
yses for daily use of 0.625-mg CEE plus daily use of 2.5-mg Gap time from menopause to first use of HT explains the medroxyprogesterone acetate from the WHI estrogen plus residual discrepancy, with hazard ratios in the observational progestin trial (14, 15) and the corresponding observational study estimated to be only 7 percent higher than those in the study subset among women with a uterus at WHI enroll- clinical trial (p ¼ 0.82) following additional control for gap ment. These analyses mutually reinforce those given here concerning gap time as a useful explanatory factor.
Among women who initiate CEE use soon after meno- The present analyses suggest a possibly reduced breast pause (e.g., <5 years), the women most likely to be making cancer risk among women who initiate CEE some years hormone therapy decisions in the future, WHI data do not (e.g., >5 years) following menopause. Although the bio- provide clear evidence for either an overall reduction or an logic basis for any such reduction is unclear, preclinical overall increase in breast cancer risk with CEE use (tables 5 studies indicate that breast cancers, when exposed to a and 6). Our interaction analyses suggest a relatively higher period of estrogen deprivation, make adaptive changes hazard ratio among women having such characteristics as low (16, 17) that alter their susceptibility to proliferative stimu- body mass index or high Gail model (6) breast cancer risk.
lation by estrogen. In addition, lobular involution is associ- The clinical trial included very few women without prior ated with reduced breast cancer risk (18), and a longer time HT and with short gap times. Hence, the hazard ratios shown from menopause with resultant involution could decrease in table 4 are not robust to gap time cutpoint choices (e.g., the number of epithelial breast cells potentially influenced 5 vs. 10 years) or other analytic choices. Even when clinical In summary, with careful standardization and control, and California at Davis, Sacramento, California: John Robbins; with consideration of time from menopause to CEE initia- University of California at Irvine, Irvine, California: F. Allan tion and time since CEE initiation, the hazard ratios from Hubbell; University of California at Los Angeles, Los Angeles, the WHI trial and cohort study agree concerning the breast California: Howard Judd; University of California at San cancer effects of CEE. Among hysterectomized women who Diego, LaJolla/Chula Vista, California: Robert D. Langer; initiate a daily 0.625 CEE regimen soon after menopause, University of Cincinnati, Cincinnati, Ohio: Margery Gass; there is little indication of a reduction in breast cancer risk.
University of Florida, Gainesville/Jacksonville, Florida:Marian Limacher; University of Hawaii, Honolulu, Hawaii:David Curb; University of Iowa, Iowa City/Davenport, Iowa:Robert Wallace; University of Massachusetts/Fallon Clinic, Worcester, Masssachusetts: Judith Ockene; University ofMedicine and Dentistry of New Jersey, Newark, New Jersey: The WHI program is supported by contracts from the Norman Lasser; University of Miami, Miami, Florida: Mary National Heart, Lung, and Blood Institute. Dr. Prentice’s Jo O’Sullivan; University of Minnesota, Minneapolis, work was partially supported by grant CA53996 from the Minnesota: Karen Margolis; University of Nevada, Reno, Nevada: Robert Brunner; University of North Carolina, The authors thank the WHI investigators and staff for Chapel Hill, North Carolina: Gerardo Heiss; University of their outstanding dedication and commitment.
Pittsburgh, Pittsburgh, Pennsylvania: Lewis Kuller; Univer- A list of key investigators involved in this research sity of Tennessee, Memphis, Tennessee: Karen C. Johnson; follows. A full listing of WHI investigators can be found at University of Texas Health Science Center, San Antonio, Texas: Robert Brzyski; University of Wisconsin, Madison, Program Office—National Heart, Lung, and Blood In- Wisconsin: Gloria E. Sarto; Wake Forest University School of stitute, Bethesda, Maryland: Elizabeth Nabel, Jacques Medicine, Winston-Salem, North Carolina: Denise Bonds; Rossouw, Shari Ludlam, Linda Pottern, Joan McGowan, and Wayne State University School of Medicine/Hutzel Leslie Ford, and Nancy Geller. Clinical Coordinating Hospital, Detroit, Michigan: Susan Hendrix.
Center—Fred Hutchinson Cancer Research Center, Seattle, Wyeth Pharmaceuticals (Madison, New Jersey) provided Washington: Ross Prentice, Garnet Anderson, Andrea medication tested in this study. Dr. Langer has been LaCroix, Charles L. Kooperberg, Ruth E. Patterson, and a consultant for Wyeth Pharmaceuticals and has received Anne McTiernan; Wake Forest University School of Med- research support from this company within the past 3 years.
icine, Winston-Salem, North Carolina: Sally Shumaker; Dr. Chlebowski is a consultant for Astra-Zeneca Pharma- Medical Research Labs, Highland Heights, Kentucky: Evan ceuticals LP (Wilmington, Delaware), Novartis (Basel, Stein; and University of California at San Francisco, San Switzerland), Pfizer Inc. (New York, New York), Eli Lilly Francisco, California: Steven Cummings. Clinical Centers— and Co. (Indianapolis, Indiana), and Organon International Albert Einstein College of Medicine, Bronx, New York: Sylvia (Kenilworth, New Jersey) and has received research support Wassertheil-Smoller; Baylor College of Medicine, Houston, from Eli Lilly and Co. and Organon. Dr. McTiernan has Texas: Jennifer Hays; Brigham and Women’s Hospital, received speaker and research support from Wyeth Phar- Harvard Medical School, Boston, Massachusetts: JoAnn maceuticals and Besins International (Paris, France) and Manson; Brown University, Providence, Rhode Island: has consulted for Novartis, Proctor & Gamble (Cincinnati, Annlouise R. Assaf; Emory University, Atlanta, Georgia: Ohio), Zymogenetics Inc. (Seattle, Washington), and Pfizer Lawrence Phillips; Fred Hutchinson Cancer Research Inc. Dr. Prentice received an honorarium from Wyeth Center, Seattle, Washington: Shirley Beresford; George Washington University Medical Center, Washington, DC:Judith Hsia; Los Angeles Biomedical Research Institute atHarbor-UCLA Medical Center, Torrance, Callifornia:Rowan Chlebowski; Kaiser Permanente Center for Health Research, Portland, Oregon: Evelyn Whitlock; Kaiser 1. Anderson GL, Limacher M, Assaf AR, et al. Effects of con- Permanente Division of Research, Oakland, California: jugated equine estrogen in postmenopausal women with Bette Caan; Medical College of Wisconsin, Milwaukee, hysterectomy: the Women’s Health Initiative randomized Wisconsin: Jane Morley Kotchen; MedStar Research controlled trial. JAMA 2004;291:1701–12.
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Source: http://kooperberg.fhcrc.org/papers/2008prentice2.pdf