Affiliations of authors: D. Wartenberg, Environmental and Occupational Health Sciences Institute, University of Medicine and Dentistry of New JerseyRobert Wood Johnson Medical School, Piscataway, NJ; E. E. Calle, M. J. Thun, C. W. Heath, Jr., C. Lally, American Cancer Society, Atlanta, GA; T. Woodruff, Office of Policy, U.S. Environmental Protection Agency, Washington, DC.
Correspondence to: Daniel Wartenberg, Ph.D., Environmental and Occupational Health Sciences Institute, 170 Frelinghuysen Rd., Piscataway, NJ 08854 (e-mail: dew{at}eohsi.rutgers.edu).
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ABSTRACT |
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INTRODUCTION |
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A second possible explanation for the different effects of ETS and mainstream smoke on breast cancer risk is that the age at first exposure may modify the effect of exposure on breast cancer risk. Women may be at greatest risk from both ETS and mainstream smoke early in life, when the breast tissue is growing most rapidly. Failure to consider age at first exposure may obscure an association with breast cancer, particularly in the subgroup of never-smoking women exposed during adolescence (1921). Most studies of ETS and mainstream smoke do not assess this potential effect explicitly. However, Calle et al. (22) found that women who started smoking before age 16 years were 1.6 (95% confidence interval [CI] = 1.22.2) times more likely to die of breast cancer than nonsmokers, whereas the risk for women who started smoking at age 20 years or older was equal to that for nonsmokers. Investigators who fail to look separately at this young age group in active or passive smokers may miss the most sensitive subgroup. Some other researchers (23,24) have reported increased breast cancer risk among women who started smoking during adolescence or prior to their first pregnancy, while still others (13,25) have reported no increase in this risk. Differences in the age at first exposure in the various study populations could explain some of the differences among study results.
A third possible explanation is that genetic polymorphisms may modify the effect of smoking on breast cancer. Ambrosone et al. (26) reported that only those postmenopausal women with the slow acetylation phenotype of the polymorphic N-acetyltransferase 2 (NAT2) polymorphism showed an association between active smoking and incident breast cancer risk. Fast acetylators showed no such association. A similar evaluation of women from the Nurses' Health Study (27), in which 76% of the subjects were postmenopausal, found a nonstatistically significantly elevated breast cancer risk of 1.5 (95% CI = 0.73.2) among slow acetylators who currently smoked at least 15 cigarettes per day as compared with never-smoking rapid acetylators. Rapid acetylators who smoke this number of cigarettes had a smaller risk of 1.2 (95% CI = 0.53.3) times that of never-smoking rapid acetylators. Morabia et al. (28) reported that, among postmenopausal slow acetylators, active smokers had a breast cancer risk of 2.5 (95% CI = 1.06.3) times that of never smokers; among postmenopausal rapid acetylators, the relative risk was 1.3 (95% CI = 0.53.2). Among premenopausal women, the relative risk of breast cancer for active smokers as compared with never smokers was 0.9 (95% CI = 0.42.2) among slow acetylators and 1.6 (95% CI = 0.64.4) among rapid acetylators. However, polymorphisms of NAT2 activity have not been reported with respect to ETS exposure and menopausal status. Because most studies of active and passive smokers were not stratified on menopausal status and NAT2 genotype, we do not know whether the populations studied were similar with respect to these two factors that seem to predict breast cancer risk and whether the comparisons are interpretable meaningfully.
Finally, tobacco-specific nitrosamines and certain other carcinogens are more concentrated in ETS than in mainstream smoke (1,2,29). Therefore, it is possible that exposure to ETS confers greater risk than active smoking alone (on a per weight basis). In addition, a greater proportion of sidestream smoke than mainstream smoke is in the vapor phase than in the particulate phase (30), resulting in greater absorption of chemicals into the blood and lymph systems (31).
To further investigate the risk of breast cancer among women exposed to ETS, we examined the association of breast cancer mortality and ETS exposure in a large, prospective cohort of women enrolled in the American Cancer Society's Cancer Prevention Study II (CPS-II). This cohort has provided a wealth of data about active smoking (29,32,33) and has been used to investigate the association between ETS exposure and mortality from lung cancer (4) and coronary heart disease (5).
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SUBJECTS AND METHODS |
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Women in this study were selected from the 676 306 female participants in CPS-II, a prospective mortality study of about 1.2 million U.S. men and women begun by the American Cancer Society in 1982 (22,34). The cohort is a convenience sample identified and enrolled by more than 77 000 volunteers in all 50 states, the District of Columbia, and Puerto Rico. Families were eligible for enrollment if at least one household member was 45 years old or older. All enrolled members were at least 30 years old. The median age of the female study participants in 1982 was 56 years; 75% of the women were between the ages of 45 and 70 years. Enrollees completed a mailed confidential questionnaire that included questions about personal identifiers: demographic characteristics; personal and family history of cancer and other diseases; and various behavioral, environmental, occupational, dietary, and (for women) reproductive exposures.
The vital status of study participants was determined from the month of enrollment through December 31, 1994, using two approaches. First, volunteers made personal inquiries in September 1984, 1986, and 1988 to determine whether enrollees were alive or deceased and to record the date and place of all deaths. Second, automated linkage using the National Death Index was used to extend follow-up through December 31, 1994 (35), and to identify deaths among 13 219 (2%) women who were lost to follow-up between 1982 and 1988. At the end of mortality follow-up in December 1994, 587 855 (86.9%) women were alive, 86 374 (12.8%) had died, and 2077 (0.3%) had had follow-up truncated on September 1, 1988, because of insufficient data for the National Death Index linkage. Death certificates or multiple causes of death codes were obtained for 98.0% of all women known to have died.
Exposure to ETS was defined in two ways. The primary analyses defined exposure to ETS as active smoking by the spouse of a never-smoking female study participant, as reported by the spouse in his questionnaire. These analyses examined current and former spousal smoking of cigarettes, cigars, and pipes and considered both amount and duration of spousal smoking. A second definition of ETS exposure was derived directly from each woman's report of the number of hours per day that she was "exposed to the smoke of others" at home, at work, and elsewhere.
Breast cancer deaths were defined as deaths through December 31, 1994, from breast cancer (International Classification of Diseases, 9th Revision, codes 174.0174.9) (36) as the underlying cause. We excluded from all analyses women who reported prevalent cancer at baseline (except nonmelanoma skin cancer), women who had ever smoked or had an unknown smoking history, women who had no spouse in the study cohort or had been married more than once, or women who had a spouse with an unknown smoking history (Table 2). In addition, for the analysis of amount and duration of exposure to spousal ETS, we also excluded women whose spouse reported incomplete information regarding amount and duration of smoking and women whose age at the time of marriage was unknown (Table 2
). For the analysis of self-reported exposure to the smoke of others, we excluded women with invalid data for these questions (Table 2
). Invalid data were responses that were omitted or incomplete, making it impossible to determine smoking status, duration, or amount.
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We used Cox proportional hazards modeling (37) to compute rate ratios (RRs) and to adjust for potential risk factors other than ETS. Cox models were adjusted for year of age at study entry, race (white, black, or other), years of education (<12, 12, 1315, or 16 years), history of breast cancer in mother or sister (yes or no), personal history of breast cysts (yes or no), age at first live birth (<20, 2024, 2529, or
30 years), age at menarche (<12, 12, 13, or
14 years), age at menopause (<45, 4549, 5054, or
55 years or premenopausal), number of spontaneous abortions (0, 1, or
2), oral contraceptive use (ever or never), use of estrogen replacement therapy (ever or never), body mass index (weight in kg/height in m2; >19.1, 19.121.9 22.027.2, 27.332.2, or
32.3), alcohol intake (none, three drinks per week to fewer than one drink per day, one drink per day, or more than one drink per day), total fat consumption (estimated grams per week categorized into quintiles) (38), vegetable consumption (frequency per week categorized into quintiles), occupation (blue collar, white collar, housewife, or unknown), and spousal occupation (blue collar, white collar, househusband, or unknown).
To evaluate the possibility of overfitting the model, we calculated both the RRs adjusted for age only and the RRs adjusted for all covariates. RRs reported in the text are from multivariate models with all variables included. To evaluate potential exposureresponse relationships, we included ordinal variables for packs per day smoked by the spouse (<1, 1, >1 to <2, or 2), number of years the spouse smoked (110, 1120, 2130, or >30 years), and pack-years smoked (112, >1225, >2541, or >41 pack-years) by the spouse and used the Wald chi-square test to test for statistical significance of a linear trend (39). Pack-years were calculated by multiplying the number of packs of cigarettes smoked per day by the number of years of smoking.
To evaluate the potential effect modification, we included statistical interaction terms between age at baseline or age at marriage and the main effect (any, current, or former smoking of her spouse) and used the chi-square test to evaluate statistical significance.
To summarize the set of published passive smoking and breast cancer studies, we conducted a simple meta-analysis (40). We assessed the homogeneity of the studies using a chi-squared test. We stratified the studies by design and calculated the average risks for each stratum under a random-effects model.
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RESULTS |
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DISCUSSION |
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The strengths of our study are its prospective design, its size, and its use of spousal rather than proxy exposure estimates. Our analyses of spousal smoking were based on 396 breast cancer deaths among ETS-exposed, never-smoking women. Only one other study (12) included more than 150 exposed case subjects, and that was a casecontrol study (Table 1). We consider exposure to spousal ETS to be a valid indicator of potential long-term exposure to ETS. For example, exposure to spousal smoking was associated with increased mortality from lung cancer and ischemic heart disease in this cohort (4,5). Our information on spousal smoking was classified into current or former exposure and by amount and duration and was examined in women who, at the time of enrollment, had married only once. Although we observed a few elevated RRs, we believe these to be spurious because of the small size of the elevations, their inconsistency across population and exposure subgroups, and the large number of statistical comparisons considered in this study.
One limitation of our study is the reliance on breast cancer mortality rather than on incidence to identify disease. Breast cancer cases that progress to death are a small subset of all cases (41). Thus, our results could reflect the potential effects of ETS on breast cancer incidence, survival, or both. Most previous studies that have found substantial breast cancer risks associated with ETS have been studies of incident breast cancers (Table 1). However, it seems unlikely that the use of mortality as an end point can explain the difference between the results of this and previous studies. If, as previous studies have suggested, never-smoking women exposed to ETS truly have a doubled risk of developing breast cancer, then, for the observed mortality effect to be null, the survival rate among women with breast cancer would have to be twice as high among women who had been exposed to ETS as among women who had not been exposed to ETS. Although this survival difference may be theoretically possible, it seems highly unlikely.
A second limitation of our study lies in our assessment of exposure to ETS. We obtained complete lifetime spousal smoking histories at time of enrollment, and these reported exposures of active smoking have been shown to be highly accurate predictors of smoking-related mortality (29,32,33). However, spousal smoking was assessed only once, at baseline, and it could have been altered over the 12-year follow-up period through changes in behavior or death. A decrease in spousal smoking, which is more likely than an increase, would have biased the results toward an apparent lack of effect of ETS on breast cancer mortality. Nonetheless, we believe a woman's long-term past exposure to spousal smoking was likely to be well captured in this study and in other studies that, using the same cohort, found positive associations between exposure to ETS and lung cancer (4) and coronary heart disease (5).
By contrast, the measure of perceived ETS exposure in the home and outside the home at the time of enrollment was a crude, self-reported measure that was the only source of information on ETS exposure from household members other than the spouse and from co-workers and others outside the house. It is possible that our lack of data on past ETS exposures outside the home (or nonspousal exposures in the home) could introduce sufficient misclassification to obscure a small but true association between breast cancer and lower levels of ETS exposure. However, cumulative exposure to spousal smoking has been found to be associated with ETS exposure outside the home, and our analyses suggest that breast cancer death rates do not increase with cumulative exposure to spousal smoking (14,16). In addition, the measurement of exposure to ETS among never-smoking women in this cohort has been sensitive enough to reveal small positive associations with mortality in previous studies (4,5).
In contrast to this study, previous studies of the association of ETS exposure and female breast cancer have shown fairly consistent positive results (Table 1). For example, using data from Hirayama's 16-year census population-based prospective mortality follow-up study of men and women aged 40 years and above (14), the study most similar to this one, Wells (16) found that the risk of breast cancer among nonsmoking women married to smokers was 1.3 (95% CI = 0.82.0) times that of nonsmoking women married to nonsmokers. In the only other cohort study of never-smoking women, Jee et al. (15) found a similarly small excess breast cancer incidence among never-smoking women married to former and current smokers as compared with never-smoking women married to nonsmokers.
Five casecontrol studies have also yielded positive results. For example, Sandler et al. (6,7) conducted a hospital-based, casecontrol study of cancer among women between the ages of 15 and 59 years, somewhat younger than CPS-II women, and found a relative risk of 2.0 (95% CI = 0.94.3) for breast cancer among those with spousal ETS exposure as compared with women married to nonsmokers, increasing from 2.03.3 as the number of household members who smoked increased (7,16). A large, casecontrol study by Smith et al. (8), which was limited to women diagnosed with breast cancer prior to age 36 years, reported a modest risk associated with ETS exposure (odds ratio [OR] = 1.6; 95% CI = 0.83.1).
In an even larger population-based, casecontrol study in Switzerland, Morabia et al. (9) found an elevated risk of breast cancer incidence among nonactive women smokers (i.e., those who smoked <100 cigarettes in their lifetime) ever exposed to ETS (OR = 3.1; 95% CI = 1.66.1) as compared with those never exposed to ETS and for those women ever married to an active smoker (OR = 2.0; 95% CI = 1.13.7) as compared with those never married to an active smoker.
Lash and Aschengrau (13), in a casecontrol study of women diagnosed with breast cancer in the mid-1980s, found that passive smokers were twice as likely to be diagnosed with breast cancer as women who had not been exposed to ETS (OR = 2.0; 95% CI = 1.13.7). Women who were first exposed to ETS before age 12 years had an OR of 4.5 (95% CI = 1.216.0), and those exposed at or after age 12 years were at smaller but still elevated risk.
Finally, Johnson et al. (1012) conducted a casecontrol study using the Canadian National Enhanced Cancer Surveillance System, mailing a questionnaire to nearly 4000 women, more than 60% of whom responded. Premenopausal never-smoking women regularly exposed to ETS had an OR 2.3 times that of never-smoking women who were not exposed to ETS (95% CI = 1.24.6) (12), and postmenopausal never-smoking women regularly exposed to ETS had an OR of 1.2 (95% CI = 0.81.8) (11,12). However, this study had several limitations, including an extremely small referent group (14 premenopausal and 52 postmenopausal case subjects), recall bias, and response bias.
Limitations applicable to most of these studies include the small number of subjects in at least some categories, possible recall bias in casecontrol studies, failure to use a consistently unexposed group for the referent, uncertainty in exposure quantification, possible uncontrolled confounding, and failure to adjust for genetic susceptibility. In addition, the age of the subjects varied substantially, complicating the consideration of the possible role of menopausal status and exposure at young ages. Many of these limitations also apply to this study.
Wells (18) reported that, when combined in a simple meta-analysis, the average relative risk of the first four studies described above [Sandler et al. (6,7), Smith et al. (8), and Morabia et al. (9)] is 1.8 (95% CI = 1.42.4). Considering all eight studies (Table 1), we find the studies to be heterogeneous (P = .011). Therefore, we stratify by study design and find each group to be statistically homogeneous (cohort studies, P = .266; casecontrol studies, P = .218). The average risk for the casecontrol studies is 1.8 (95% CI = 1.42.5), whereas that for the cohort studies is 1.1 (95% CI = 0.91.4).
The disparity between the results for casecontrol and cohort studies is the most perplexing observation of our study. One possible explanation is that the casecontrol studies are biased. Typical biases found in casecontrol studies include recall bias and selection bias. However, the consistency of results among the casecontrol studies suggests that selection bias or other design-related issues are unlikely explanations for the higher risk found in the casecontrol studies.
The cohort studies, although prospective, had relatively short follow-up times of 16 years (14), 4 years (15), and 12 years (this study). Moreover, in each of these studies, reported smoking history reflected lifetime exposures, with inadequate recall resulting in possible misclassification. Assuming that this misclassification among healthy subjects occurred at the same rate for those exposed to ETS and those unexposed, it likely would lead to an apparent lack of effect of ETS on breast cancer mortality. However, in each of these studies, other smoking-related cancers were associated with ETS exposure among nonsmoking wives, arguing against the existence of substantial misclassification.
Another difference between the casecontrol and cohort studies is that all of the casecontrol studies used the incidence of breast cancer as their study end point, but two of three cohort studies used breast cancer mortality as their end point. Although we argued earlier in this article that we do not believe that this difference in the end point studied can explain the disparity in the results, we cannot rule it out completely.
Clarification of several issues would help us better understand the differences in the reported associations between breast cancer mortality in never-smoking women and ETS exposure. First, development of reliable, validated smoking histories would be helpful in characterizing exposures. Although we believe that changes in spousal smoking are unlikely to explain the lack of observed association between ETS exposure and breast cancer mortality among never-smoking women, particularly in the cohort studies, data to support this view would be helpful. Second, assessment of possible genetic-effect modification would help clarify whether variations in genetic makeup is a possible explanation for disparate findings. Third, the roles of adolescent exposures and menopausal status deserve further scrutiny. Although most studies adjusted for age, only a few casecontrol studies were stratified on either adolescent exposure or menopausal status. Of interest, stratified results for genetic interactions were stronger in postmenopausal women (26,28), whereas passive smoking exposure effects were stronger in premenopausal women (12,16).
In conclusion, given the importance of understanding the many causes of female breast cancer, the potentially high population-attributable risk of ETS exposure (even if the relative risk is modest), and the limited number of previous studies of this issue, our analyses provide an important contribution to the study of the association between exposure to ETS and female breast cancer.
Overall, the results are null. That is, we see little evidence of an association between exposure to ETS and the risk of dying of breast cancer. These results contradict results from previous studies, which were mainly casecontrol studies, but they are compelling in that they come from a large prospective cohort study. We did find data that suggest that women exposed to ETS under age 20 years may be at increased risk of dying of breast cancer, however, and this observation warrants follow-up.
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NOTES |
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Supported by U.S. Environmental Protection Agency grant 992097-1 and by Public Health Service grant ES0502209 from the National Institute of Environmental Health Sciences, National Institutes of Health, Department of Health and Human Services.
We thank Alfredo Morabia for providing helpful comments on the manuscript.
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Manuscript received January 4, 2000; revised August 14, 2000; accepted August 16, 2000.
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