REPORTS

Estrogen Replacement Therapy and Breast Cancer Survival in a Large Screening Study

Catherine Schairer, Mitchell Gail, Celia Byrne, Philip S. Rosenberg, Susan R. Sturgeon, Louise A. Brinton, Robert N. Hoover

Affiliations of authors: C. Schairer, S. R. Sturgeon, L. A. Brinton, R. N. Hoover (Environmental Epidemiology Branch), M. Gail, P. S. Rosenberg (Biostatistics Branch), Division of Cancer Epidemiology and Genetics, National Cancer Institute, Bethesda, MD; C. Byrne, Channing Laboratory, Harvard Medical School, Boston, MA.

Correspondence to: Catherine Schairer, Ph.D., National Institutes of Health, EPN, Rm. 443, 6130 Executive Blvd., MSC 7374, Bethesda, MD 20892-7374 (e-mail: Schairec{at}epndce.nci.nih.gov).


    ABSTRACT
 Top
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
BACKGROUND: Hormone replacement therapy has been associated in some studies with reductions in breast cancer mortality among women who develop this disease. It is unclear whether this association reflects the biologic activity of the hormones or the earlier detection of tumors among hormone users. We examined breast cancer mortality among women who were diagnosed with axillary lymph node-negative and node-positive breast cancer according to the currency of estrogen use at diagnosis. METHODS: Vital status through June 1995 was determined for 2614 patients with postmenopausal breast cancer diagnosed during the period from 1973 to January 1981. We estimated adjusted hazard-rate ratios (adjusting for tumor size, age, race, Quetelet [body mass] index, and number of positive lymph nodes in women with node-positive disease) and unadjusted cumulative probabilities of breast cancer death over time since diagnosis. RESULTS: Among patients with node-negative disease, rate ratios for breast cancer mortality associated with current use compared with nonuse at diagnosis were 0.5 (95% confidence interval [CI] = 0.3-0.8) until 144 months after diagnosis and 2.2 (95% CI = 0.9-5.2) thereafter. Mortality was not statistically significantly lower in past users. The cumulative probabilities of breast cancer mortality at the end of follow-up were 0.14, 0.14, and 0.09 in nonusers, past users, and current users, respectively. Among women with node-positive disease, the rate ratios associated with current and past use were both 0.5 until 48 months after diagnosis (95% CI = 0.3-0.8 for current users; 95% CI = 0.3-0.9 for past users) and were 1.1 (95% CI = 0.7-1.7) and 1.8 (95% CI = 1.2-2.7), respectively, thereafter. The cumulative probabilities of breast cancer mortality were 0.32, 0.39, and 0.27 in nonusers, past users, and current users, respectively. CONCLUSIONS: Patients with breast cancer who were using replacement estrogens at the time of diagnosis experienced reductions in breast cancer mortality, which waned with the time since diagnosis.



    INTRODUCTION
 Top
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
Although hormone replacement therapy is associated with an increased incidence of breast cancer (1), it has been associated with lower mortality from breast cancer or improved survival after a diagnosis of breast cancer in some (2-11), but not all (12-16), studies. The observation that hormones preferentially increase the risk of developing less advanced tumors (1) raises the possibility that the lower mortality in hormone users reflects increased breast cancer surveillance, rather than a biologic effect of hormones. Clarification of the role of exogenous hormones on breast cancer prognosis is important both for further understanding of the role of hormones in the carcinogenic process and for weighing the risks and benefits of hormone replacement therapy (17).

To address further the relationship between hormone replacement therapy and breast cancer survival, we examined death from breast cancer according to currency of hormone use at diagnosis among 2614 breast cancer cases diagnosed during the period from 1973 to January 1981 and followed for vital status through June 1995.


    SUBJECTS AND METHODS
 Top
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
Study Subjects

Study subjects were participants in the Breast Cancer Detection Demonstration Project (BCDDP), a 5-year breast cancer screening program sponsored by the National Cancer Institute and the American Cancer Society. Conducted during the period from 1973 through 1980, the BCDDP involved more than 280 000 women at 29 centers in 27 cities throughout the United States. Five annual examinations, each including a physical examination of the breast and mammography, were offered to each screening participant. A total of 4363 breast cancers were detected among the screening participants during the 5-year course of the project.

In 1980, at the end of the BCDDP, the National Cancer Institute initiated a follow-up study that included, among others, all patients with breast cancer that had been diagnosed during the BCDDP. The follow-up study included yearly telephone interviews until 1986 and continued with two self-administered, mailed questionnaires, the first during the period from 1987 through 1989 and the second during the period from 1993 through June 1995. We made vigorous efforts to locate and obtain the vital status of all participants in the follow-up study, which included a search of the National Death Index through 1993. The follow-up study was approved by the Institutional Review Board at the National Cancer Institute.

Data

We limited our analysis to women whose first diagnosis of breast cancer occurred during the BCDDP and who were menopausal at the time of diagnosis (n = 2675). We ascertained vital status through the last mailing of questionnaires (in 1995). A total of 1039 women died during the follow-up period from 1974 through June 1995. For the 1636 women who did not die during the follow-up period, we ascertained vital status for 2.6% through 1992, for 34.8% through 1993, for 57.5% through 1994, and for 5.2% into 1995. We obtained death certificates for 97.7% of the identified deaths, allowing us to code cause of death. We excluded 61 subjects whose death certificates were not available, leaving an analytic dataset of 2614 women. A total of 486 deaths from breast cancer and 492 deaths from causes other than breast cancer were identified during follow-up among these study subjects.

Data Sources

We obtained information on female hormone use and other breast cancer risk factors from the following sources: 1) questionnaires completed at each of the five annual screening visits of the BCDDP; 2) an in-home interview administered as part of a case-control study involving breast cancer cases diagnosed during the BCDDP, which was completed by 72.3% of the 2614 members of the analytic cohort (18); and 3) telephone interviews administered as part of the follow-up study, which were completed by 87% of the analytic cohort.

Exposure Definitions

Information on the noncontraceptive use of female hormones was available from all three data sources listed above. Only the case-control interview, however, elicited details on the type, name, and dose of hormones used in relationship to menopause. Among patients with breast cancer included in the case-control study who used hormones, 1.25 mg of Premarin was the preparation used longest by 40%, 0.625 mg of Premarin was the preparation used longest by 17%, other doses of Premarin were the preparations used longest by 15%, diethylstilbestrol was the preparation used longest by 4%, and other hormones were the preparations used longest by 9%. Fourteen percent of patients who used hormones were uncertain of the type of hormone used longest (18). Progestins were not prescribed widely in the 1970s when these women were diagnosed with breast cancer.

Information on current, past, or never use of hormones at the time of breast cancer diagnosis was available for 94.7% of the study subjects. For 76.4% of the subjects, the currency of hormone use was based on the BCDDP screening form completed within a year before breast cancer diagnosis. Some subjects were administered versions of these forms that ascertained currency of hormone use but did not distinguish between past use and never use of hormones. For the 14.1% of study subjects who reported on one of these forms that they were not currently using hormones, information from the casecontrol and follow-up interviews was used to separate past users from never users. Similarly, for the 9.5% of study subjects for whom currency of use was unknown on the basis of the screening forms (for instance, some subjects did not have a screening form within a year of diagnosis), information from the available case-control and follow-up interviews was used to determine currency of use. Status of hormone use was not defined for 61 women who died of breast cancer (12.6% of breast cancer case patients), for 31 women who died of other causes (6.3% of other causes of death), and for 48 women who did not die during the follow-up period (2.9% of surviving case patients).

Information on duration of hormone use was available from the case-control or follow-up interview for 65% of hormone users. Information on time since last use was available for 76% of past users.

We determined menopausal status at the time of breast cancer diagnosis from information obtained at the case-control interview, from the baseline interview of the follow-up study, or from the most recent screening form that was obtained prior to the diagnosis of breast cancer (in that order of preference). This hierarchical scheme was used because the case-control interview provided the most detailed information on menopausal status and the screening forms provided the least. The following women were included in the analysis if their age at diagnosis was at least 1 year greater than the age at their last reported period: 1) women who stopped menstruating because of removal of both ovaries, 2) women who stopped menstruating because of a hysterectomy if their age at surgery was greater than the median age at natural menopause in this cohort (51.1 years), 3) women who reported a natural menopause (defined as not having a period for 3 months), and 4) the 281 women (11%) without a case-control or follow-up interview if they reported no longer having regular periods on the closest screening form prior to their date of diagnosis.

Information on tumor size and axillary lymph node status was available from standardized pathology reports obtained for 95% of study subjects as part of the BCDDP (19). No information was available on presence of distant metastases. We designated extent of disease as follows: in situ (both lobular and ductal); lymph node-negative with tumor sizes of less than 2 cm, 2 cm or more and less than 5 cm, or 5 cm or more; and lymph node-positive with tumor sizes of less than 2 cm, 2 cm or more and less than 5 cm, or 5 cm or more.

We classified the histology of invasive tumors into the following five categories: 1) carcinoma of the breast with productive fibrosis, carcinoma of the breast (not otherwise specified), metaplastic mammary carcinoma, or minimally invasive infiltrating duct cell carcinomas; 2) medullary or lobular carcinoma; 3) comedo or papillary carcinoma; 4) mucinous or tubular carcinoma; and 5) unknown histology (20). For cases with multiple histologies, the lowest numbered category was used.

We obtained demographic information from forms filled out upon entry into the screening program, and we obtained information on height and weight from the form completed at the closest screening visit before the diagnosis of breast cancer. The Quetelet (body mass) index was calculated as weight in kilograms divided by height in square meters.

Statistical Analyses

Death from breast cancer, determined from the underlying cause of death on the death certificate [International Classification of Diseases for Oncology codes 1748 or 1749 (21)], was the primary outcome of interest. We considered deaths from causes other than breast cancer to be competing risks. We censored survival time at the date of death or the date of last interview or contact.

We used Cox proportional hazards models to obtain univariate and multivariate hazard rate ratio estimates of death from breast cancer associated with hormone use and other factors (22). The time scale was months after a breast cancer diagnosis. We assessed the appropriateness of the proportional hazards model by testing the significance of an interaction term between covariates and time in the proportional hazards model. Because we found significant evidence of nonproportionality of the hazards for groups defined by age at diagnosis, we stratified by age at diagnosis in the proportional hazards models. Hormone use, the primary exposure of interest, also did not satisfy the proportional hazards assumption over the entire period of follow-up. Therefore, we calculated hazard rate ratios for different time periods since diagnosis to evaluate the effects of hormone exposure.

To determine proportions of study subjects dying of breast cancer according to hormone use, we calculated cumulative crude probabilities of death from breast cancer by time t in the presence of competing risks, unadjusted for other prognostic factors (23). The statistical significance of the differences in the cumulative probability among exposure groups was derived from statistics based on cumulative weighted differences, as in the paper by Pepe and Mori (23). These probabilities are observable quantities that reflect the effect of an exposure on competing causes of death as well as the cause of death of interest. To put confidence intervals (CIs) on estimated cumulative crude probabilities of death from breast cancer, we applied counting process methods to calculate the variance of quantities UjPR, defined in the paper by Pepe and Mori (23).

We calculated the median time to death among those who had died of breast cancer by the end of the study period from the distribution of the ratio of the cumulative probability of dying of breast cancer by time t and the cumulative probability of dying of breast cancer at the end of the follow-up period (24).

We used software developed by Pepe and Mori (23) to calculate crude probabilities. We used standard chi-squared tests to evaluate associations in contingency tables, and we used the notation {chi}2 (df) to denote a chi-squared distribution with degrees of freedom (df) (25). All P values are two-sided and were considered statistically significant for P<.05.


    RESULTS
 Top
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
Subjects were followed, on average, for 14.1 years after a diagnosis of breast cancer. The median follow-up time was 16.0 years, with a maximum of 21.3 years and a minimum of less than a year. A total of 978 deaths occurred among the 2614 study subjects, including 486 deaths from breast cancer.

Prognostic Factors

Table 1Go presents hazard ratios for potential prognostic factors for breast cancer survival, adjusted for other factors in the table. As expected, the extent of disease at diagnosis was a very strong prognostic factor, as was the number of positive lymph nodes in lymph node-positive women. Patients with mucinous or tubular invasive tumors were at lower risk than those with invasive tumors of other histologic types. Breast cancer mortality was greater in patients with a high body mass index, with those in the highest compared with the lowest quartile having a 60% excess mortality. Blacks and those of other races were at 70% higher risk than whites.


View this table:
[in this window]
[in a new window]
 
Table 1. Hazard ratios (HRs) indicating effect of selected characteristics on breast cancer mortality rates following diagnosis*

 
Hazard rates associated with age were not proportional over the follow-up period; women 55 years of age and older were at reduced risk of death from breast cancer compared with younger women for the first 48 months since diagnosis but were at higher risk after 144 months since diagnosis (data not shown). Level of income, educational level, and marital status were not prognostic factors before or after adjustment for other variables.

As shown in Table 2,Go current users of hormones were more likely to have been diagnosed with in situ tumors than nonusers (P = .001, {chi}2 test, df = 18). There were no remarkable differences in other tumor characteristics according to hormone use. A higher proportion of current users had a lower body mass index than nonusers or past users (P<.0001, {chi}2 test, df = 8), were white (P = .0005, {chi}2 test, df = 8), and were younger at diagnosis (P<.0001, {chi}2 test, df = 12).


View this table:
[in this window]
[in a new window]
 
Table 2. Distribution of selected characteristics by hormone use*

 
Hazard Rates According to Hormone Use

During the follow-up period, 184 of the breast cancer deaths occurred among hormone nonusers at diagnosis, 135 among past users, and 106 among current users. We were unable to determine whether 43 women who died of breast cancer were nonusers or past users, whether 17 were current or past users, and whether one had used hormones at all. Subsequent analyses focus on the subjects who were known to be nonusers, past users, or current users.

Because the hazards associated with nonuse, past use, and current use of hormones were clearly nonproportional over the entire follow-up period and differed in women with lymph node-negative and lymph node-positive breast cancer, we present results according to lymph node status and for different periods of follow-up. Among lymph node-negative women (Table 3, A),Go including those with in situ cancer, current use of hormones compared with nonuse was associated with a 40%-60% reduction in breast cancer mortality rates during the first 144 months since diagnosis and a twofold increase thereafter after adjustment for age at diagnosis, race, Quetelet index, and tumor size. When the first 144 months of follow-up were considered as a single time period, the hazard-rate ratio associated with current use after adjustment for other factors was 0.5 (95% CI = 0.3-0.8). In past users at diagnosis, smaller reductions in breast cancer mortality were evident for the first 96 months after diagnosis; however, thereafter, there were increases that were not statistically significant. When the first 96 months was considered as a single time period, the hazard-rate ratio associated with past use was 0.7 (95% CI = 0.5-1.1).


View this table:
[in this window]
[in a new window]
 
Table 3. Hazard rate ratios (HRs) of breast cancer mortality according to hormone use and lymph node status

 
To provide further assurance that the reduction in mortality associated with current use was not due to greater breast cancer surveillance in hormone users before entering the screening program, we repeated these analyses after eliminating cases detected at the first BCDDP screening visit. Results were essentially unchanged; hazard-rate ratios for current use for the four time periods were 0.6, 0.3, 0.7, and 1.7, and those for past use for the four time periods were 0.6, 0.8, 1.3, and 1.4. Results were also similar when in situ cases were excluded and when analyses were limited to invasive cancer with tumor size less than 2 cm.

There were no remarkable differences in these results according to race or levels of Quetelet index. However, results differed according to age at diagnosis. In women less than 60 years old at diagnosis, the hazard-rate ratios for past and current use for the first 144 months since diagnosis were 0.6 (95% CI = 0.4-1.0) and 0.4 (95% CI = 0.2-0.6), respectively; in contrast, in older women, they were 1.0 (95% CI = 0.6-1.8) and 0.7 (95% CI = 0.4-1.4). The attenuation of the protective effect in older women was largely due to a decrease in the hazard rate among older nonusers of hormones, with no corresponding decrease in the hazard rate among current users.

As shown in Table 3, B,Go current and past hormone use among lymph node-positive women was associated with a 50% reduction in mortality from breast cancer in the first 48 months after diagnosis compared with hormone nonusers—but not thereafter—after adjustment for race, Quetelet index, tumor size, and number of positive lymph nodes. The estimates for past and current use for the entire period after 48 months were 1.8 (95% CI = 1.2-2.7) and 1.1 (95% CI = 0.7-1.7), respectively. Elimination of cases detected during a woman's first screening visit yielded similar results; hazard-rate ratios for current use for the four time periods were 0.4, 0.9, 1.1, and 2.0 and for past use were 0.5, 1.2, 1.3, and 3.8. Results were similar when analyses were limited to those with tumors smaller than 2 cm. Results did not differ substantially according to race, age, or levels of Quetelet index.

Among lymph node-negative women who were current users at diagnosis, duration of use was not associated with risk during the first 144 months since diagnosis; the hazard-rate ratios were 0.3 (95% CI = 0.1-0.7), 0.2 (95% CI = 0.1-0.6), and 0.4 (95% CI = 0.2-0.9) for less than 5 years, 5-9 years, and 10 or more years of use, respectively. Similarly, over the first 96 months since diagnosis, time since last use was not associated with risk among lymph node-negative past users; the hazard-rate ratios were 0.6 (95% CI = 0.3-1.2) for at least 10 years since last use, 0.4 (95% CI = 0.2-1.2) for 5-9 years since last use, and 0.6 (95% CI = 0.3-1.2) for less than 5 years since last use. The data were insufficient to effectively evaluate duration of use and time since last use in lymph node-positive women.

Probabilities of Death From Breast Cancer

In lymph node-negative women, the unadjusted cumulative crude probabilities of death from breast cancer at the end of the follow-up period in nonusers, past users, and current users of hormones were 0.14 (95% CI = 0.12-0.18), 0.14 (95% CI = 0.10-0.18), and 0.09 (95% CI = 0.07-0.12), respectively (P values for the cumulative weighted differences between nonusers and current users and nonusers and past users = .001 and .41, respectively) (Fig. 1, A).Go At 144 months since diagnosis (until which time hazard rates were lower in current users than in nonusers), the corresponding probabilities were 0.12 (95% CI = 0.10-0.15), 0.10 (95% CI = 0.08-0.13), and 0.06 (95% CI = 0.04-0.08). Ratios of these probabilities for past users compared with nonusers and current users compared with nonusers were 1.0 and 0.6, respectively, at the end of follow-up and 0.8 and 0.5 at 144 months. The median times to death from breast cancer among those who eventually died of breast cancer during the follow-up period were 66 months, 90 months, and 87 months in nonusers, past users, and current users at diagnosis, respectively.



View larger version (32K):
[in this window]
[in a new window]
 
Fig. 1. Unadjusted cumulative crude breast cancer mortality (%) according to axillary lymph node status at diagnosis for nonusers, past users, and current users of estrogen replacement therapy. P values are two-sided and statistically significant for P<.05. A) Women with lymph node-negative disease. Cumulative crude mortality for nonusers at 80, 160, and 240 months = 0.09 (95% confidence interval [CI] = 0.07-0.11), 0.12 (95% CI = 0.10-0.15), and 0.14 (95% CI = 0.12-0.18), respectively; for past users at 80, 160, and 240 months = 0.06 (95% CI = 0.05-0.09), 0.11 (95% CI = 0.09-0.14), and 0.14 (95% CI = 0.10-0.18), respectively; and for current users at 80, 160, and 240 months = 0.04 (95% CI = 0.03-0.06), 0.07 (95% CI = 0.05-0.09), and 0.09 (95% CI = 0.07-0.12), respectively. B) Women with lymph node-positive disease. Cumulative crude mortality for nonusers at 80, 160, and 240 months = 0.22 (95% CI = 0.18-0.26), 0.30 (95% CI = 0.25-0.34), and 0.32 (95% CI = 0.26-0.38), respectively; for past users at 80, 160, and 240 months = 0.18 (95% CI = 0.14-0.23), 0.32 (95% CI = 0.26-0.38), and 0.39 (95% CI = 0.31-0.47), respectively; and for current users at 80, 160, and 240 months = 0.16 (95% CI = 0.12-0.21), 0.25 (95% CI = 0.20-0.30), and 0.27 (95% CI = 0.22-0.34), respectively.

 
The unadjusted cumulative crude probabilities of death from breast cancer in lymph node-positive women for nonusers, past users, and current users were 0.32 (95% CI = 0.26-0.38), 0.39 (95% CI = 0.31-0.47), and 0.27 (95% CI = 0.22-0.34), respectively (Fig. 1, B).Go The P values for the cumulative weighted differences between nonusers and current users and nonusers and past users were .18 and .57, respectively. At 48 months after diagnosis (until which time the hazard rate was lower in current and past hormone users than in hormone nonusers), the cumulative probabilities of death were 0.16 (95% CI = 0.13-0.20) in hormone nonusers, 0.09 (95% CI = 0.06-0.14) in past users, and 0.09 (95% CI = 0.6-0.13) in current users. Ratios of crude probabilities at the end of the follow-up period were 1.2 for past users compared with nonusers and 0.8 for current users compared with nonusers. The corresponding ratios at 48 months were 0.6 and 0.6. The median times to death among those who died of breast cancer were 47, 83, and 68 months in nonusers, past users, and current users, respectively.


    DISCUSSION
 Top
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 
Breast cancer patients who had used estrogen replacement therapy experienced reductions in breast cancer mortality, which waned with time since diagnosis. These reductions in mortality were not due to differences between hormone users and nonusers in tumor size, race, age at diagnosis, Quetelet index, tumor histology, or number of positive lymph nodes. Our results are consistent with some (9-11,26,27), but not all (13-16), other studies of this topic, although others have not examined the results by lymph node status or over such a long period of follow-up. Our results are not directlycomparable to those from studies of hormone use and breast cancer mortality in women without breast cancer, because results from such studies reflect both incidence and survival (2-8,12).

The converging of the hazard rates over time since diagnosis in hormone users and nonusers may reflect selection effects due to variation in the risk of breast cancer mortality among individuals according to hormone use. Thus, a larger proportion of hormone nonusers may have died early because they had tumors with adverse prognostic factors for which we were unable to adjust, with the surviving individuals being at low risk, while a larger proportion of hormone users may have died later because they had tumors with more favorable prognostic factors (28). The longer median times to death for those who died of breast cancer among hormone users than among nonusers are consistent with this explanation. Potentially important prognostic factors for which we were unable to adjust include the degree of tumor differentiation (15,29), the aneuploidy of the tumors (16,29), and the hormone receptor status of the tumors (30). It is also possible that the delay in death from breast cancer in hormone users, who usually stopped taking hormones at the time of diagnosis, is similar to the regression sometimes seen upon stopping treatment of breast cancer with pharmacologic doses of estrogens when the disease progressed (31).

Several methodologic issues need to be considered in interpreting our results. Although the ascertainment of cases during a breast cancer screening program minimized differences in screening between hormone users and nonusers, it is still possible that differences in surveillance before or during the screening program could have confounded our results. This is particularly true among those with lymph node-positive breast cancer because we did not have information on distant metastases at diagnosis.

The absence of information on treatment is another potential concern. If current or past hormone replacement therapy is associated with the use of beneficial treatment regimens but does not itself favorably affect the carcinogenic process, our finding of a reduction in mortality would be misleading. If, however, hormone replacement therapy renders the tumor more responsive to treatment (i.e., to adjuvant endocrine therapy) or otherwise favorably affects the carcinogenic process, our basic finding of a reduction in mortality is valid. However, the magnitude of the reduction in risk due solely to the effect of the hormones is unclear.

Further research is needed to assess the extent to which reductions in breast cancer mortality associated with use of hormone replacement therapy at or prior to breast cancer diagnosis reflect biologic differences in tumors between hormone users and nonusers, an effect of hormones on tumor growth, or some other explanation, such as treatment differences. The effects on breast cancer mortality of continuing or initiating hormone replacement therapy after a breast cancer diagnosis also need to be resolved.


    REFERENCES
 Top
 Abstract
 Introduction
 Subjects and Methods
 Results
 Discussion
 References
 

1 Collaborative Group on Hormonal Factors in Breast Cancer. Breast cancer and hormone replacement therapy: collaborative reanalysis of data from 51 epidemiologic studies of 52 705 women with breast cancer and 108 411 women without breast cancer. Lancet 1997;350:1047-59.[Medline]

2 Hunt K, Vessey M, McPherson K, Coleman M. Long-term surveillance of mortality and cancer incidence in women receiving hormone replacement therapy.Br J Obstet Gynaecol 1987;94:620-35.[Medline]

3 Hunt K, Vessey M, McPherson K. Mortality in a cohort of long-term users of hormone replacement therapy: an updated analysis. Br J Obstet Gynaecol 1990;97:1080-6.[Medline]

4 Henderson BE, Paganini-Hill A, Ross RK. Decreased mortality in users of estrogen replacement therapy. Arch Intern Med 1991;151: 75-8.[Abstract]

5 Yuen J, Persson I, Bergkvist L, Hoover R, Schairer C, Adami HO. Hormone replacement therapy and breast cancer mortality in Swedish women: results after adjustment for `healthy drug-user' effect. Cancer Causes Control 1993;4:369-74.[Medline]

6 Sturgeon SR, Schairer C, Brinton LA, Pearson T, Hoover RN. Evidence of a healthy estrogen user survivor effect. Epidemiology 1995;6:227-31.[Medline]

7 Persson I, Yuen J, Bergkvist L, Schairer C. Cancer incidence and mortality in women receiving estrogen and estrogen-progestin replacement therapy—long-term follow-up of a Swedish cohort. Int J Cancer 1996;67: 327-32.[Medline]

8 Willis DB, Calle EE, Miracle-McMahill HL, Heath CW Jr. Estrogen replacement therapy and risk of fatal breast cancer in a prospective cohort of postmenopausal women in the United States. Cancer Causes Control 1996;7: 449-57.[Medline]

9 Colditz GA, Stampfer MJ, Willett WC, Hennekens CH, Rosner B, Speizer FE. Prospective study of estrogen replacement therapy and risk of breast cancer in postmenopausal women [published erratum appears in JAMA 1991;265:1828]. JAMA 1990;264: 2648-53.[Abstract]

10 Colditz GA, Stampfer MJ, Willett WC, Hunter DJ, Manson JE, Hennekens CH, et al. Type of postmenopausal hormone use and risk of breast cancer: 12-year follow-up from the Nurses' Health Study. Cancer Causes Control 1992;3:433-9.[Medline]

11 Bergkvist L, Adami HO, Persson I, Bergstrom R, Krusemo UB. Prognosis after breast cancer in women exposed to estrogen and estrogenprogestogen replacement therapy. Am J Epidemiol 1989;130:221-8.[Abstract]

12 Grodstein F, Stampfer MJ, Colditz GA, Willett WC, Manson JE, Joffe M, et al. Postmenopausal hormone therapy and mortality. N Engl J Med 1997;336:1769-75.[Abstract/Free Full Text]

13 Ewertz M, Gillanders S, Meyer L, Zedeler K. Survival of breast cancer patients in relation to factors which affect the risk of developing breast cancer. Int J Cancer 1991;49:526-30.[Medline]

14 Strickland DM, Gambrell RD Jr, Butzin CA, Strickland K. The relationship between breast cancer survival and prior postmenopausal estrogen use. Obstet Gynecol 1992; 80:400-4.[Abstract]

15 Bonnier P, Romain S, Giacalone PL, Laffargue F, Martin PM, Piana L. Clinical and biologic prognostic factors in breast cancer diagnosed during postmenopausal hormone replacement therapy. Obstet Gynecol 1995;85:11-7.[Abstract/Free Full Text]

16 Magnusson C, Holmberg L, Norden T, Lindgren A, Persson I. Prognostic characteristics in breast cancers after hormone replacement therapy. Breast Cancer Res Treat 1996;38:325-34.[Medline]

17 Col NF, Eckman MH, Karas RH, Pauker SG, Goldberg RJ, Ross EM, et al. Patient-specific decisions about hormone replacement therapy in postmenopausal women. JAMA 1997;277:1140-7.[Abstract]

18 Brinton LA, Hoover R, Fraumeni JF Jr. Menopausal oestrogens and breast cancer risk: an expanded case-control study. Br J Cancer 1986;54:825-32.[Medline]

19 Byrne C, Smart CR, Chu KC, Hartmann WH. Survival advantage differences by age. Evaluation of the extended follow-up of the Breast Cancer Detection Demonstration Project. Cancer 1994;74(1 Suppl):301-10.[Medline]

20 Tabar L, Fagerberg G, Chen HH, Duffy SW, Gad A. Tumour development, histology and grade of breast cancers: prognosis and progression. Int J Cancer 1996;66:413-9.[Medline]

21 International Classification of Diseases for Oncology. Geneva (Switzerland): World Health Organization; 1976.

22 Prentice RL, Kalbfleisch JD, Peterson AV Jr, Flournoy N, Farewell VT, Breslow NE. The analysis of failure times in the presence of competing risks. Biometrics 1978;34: 541-54.[Medline]

23 Pepe MS, Mori M. Kaplan-Meier, marginal or conditional probability curves in summarizing competing risks failure time data? Stat Med 1993;12:737-51.[Medline]

24 Gaynor JJ, Feuer EJ, Tan CC, Wu DH, Little CR, Straus DJ, et al. On the use of cause-specific failure and conditional failure probabilities: examples from clinical oncology data. J Am Stat Assoc 1993;88:400-9.

25 Snedecor GW, Cochran WG. Statistical methods. 6th ed. Ames (IA): The Iowa State University Press; 1967.

26 Burch JC, Byrd BF, Vaughn WK. Results of estrogen treatment in one thousand hysterectomized women for 14 318 years. In: van Keep PA, Greenblatt RB, Albeaux-Fernet M, editors. Consensus on Menopausal Research. Lancaster (U.K.): MTP Press Ltd.; 1976. p. 164-9.

27 Gambrell RD Jr. Proposal to decrease the risk and improve the prognosis of breast cancer. Am J Obstet Gynecol 1984;150:119-32.[Medline]

28 Aalen OO. Effects of frailty in survival analysis. Stat Methods Med Res 1994;3: 227-43.[Medline]

29 Donegan WL. Tumor-related prognostic factors for breast cancer. CA Cancer J Clin 1997;47:28-51.[Abstract/Free Full Text]

30 Donegan WL. Prognostic factors. Stage and receptor status in breast cancer. Cancer 1992;70(6 Suppl):1755-64.[Medline]

31 Nesto RW, Cady B, Oberfield RA, Pazianos AG, Salzman FA. Rebound response after estrogen therapy for metastatic breast cancer. Cancer 1976;38:1834-7.[Medline]

Manuscript received June 10, 1998; revised November 17, 1998; accepted November 27, 1998.


This article has been cited by other articles in HighWire Press-hosted journals:


             
Copyright © 1999 Oxford University Press (unless otherwise stated)
Oxford University Press Privacy Policy and Legal Statement