From the Department of Epidemiology, Graduate School of Public Health, and the University of Pittsburgh Cancer Institute, University of Pittsburgh, Pittsburgh, PA.
Received for publication April 30, 2004; accepted for publication October 14, 2004.
![]() |
ABSTRACT |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
abortion, induced; abortion, spontaneous; case-control studies; humans; ovarian neoplasms; pregnancy
![]() |
INTRODUCTION |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
It is well documented that a single livebirth compared with nulliparity significantly reduces the risk of ovarian cancer, and the risk declines 1022 percent for each additional full-term pregnancy (215). For many years, the reduced, dose-dependent risk associated with each full-term pregnancy has added to the epidemiologic evidence for two leading etiologic hypotheses of ovarian cancer. The first is the ovulation hypothesis, which posits that chronic repeated ovulation without a pregnancy-induced respite contributes to neoplasia of the ovarian epithelium (16, 17). Pregnancy is also associated with lower basal and peak gonadotropin exposure, providing biologic support for a second etiologic hypothesis implicating elevated pituitary gonadotropin secretion (4).
Although both hypotheses would predict some protection with a pregnancy of any gestational length, the extent to which ovarian cancer risk is affected by pregnancies that terminate early in gestation is unclear (18). Numerous studies have demonstrated a significant inverse relation between ovarian cancer risk and number of incomplete pregnancies (10, 1924), while others have shown no significant association between ovarian cancer risk and incomplete pregnancy (7, 9, 17, 2530). Only one group of investigators has reported a significantly increased risk of ovarian cancer in relation to history of incomplete pregnancy (31). Studies that have separated incomplete pregnancies by type (spontaneous and induced abortions) have also produced conflicting results (26, 8, 9, 13, 18, 21, 26, 29, 3239). Because these two types of interrupted pregnancies may be associated with different biologic effects, separate evaluation is needed (40, 41). In addition to differing definitions of incomplete pregnancy, interpretation of prior research is complicated by a paucity of attention to duration of incomplete pregnancies (29, 37). Hence, we undertook a study to evaluate the relations of gestational length and timing and type of incomplete pregnancy to ovarian cancer risk among women enrolled in a large population-based case-control study of ovarian cancer.
![]() |
MATERIALS AND METHODS |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
Community controls aged 65 years or younger were ascertained through random digit dialing and were frequency-matched to cases at a 1:2 ratio by 5-year age group and three-digit telephone exchange. Of the 14,551 telephone numbers screened for this purpose, 1,637 households with a potentially eligible control were identified; of these, 1,215 persons (74 percent) completed interviews. Controls aged 6569 years were ascertained through Health Care Financing Administration lists. Of the 263 potentially eligible participants identified, 152 (58 percent) were interviewed. Thus, of the 1,900 screened and potentially eligible controls, 1,367 completed control interviews (72 percent). Frequency-matched control selection was accomplished in intervals after ascertainment of each batch of cases and approximated the distribution of both telephone exchange (a surrogate measure of socioeconomic status) and age range. Interim analyses conducted during the study ensured that age was relatively equally distributed between cases and controls. However, the large number of telephone exchanges made checking for the evenness of the distribution of this variable more difficult. Nonetheless, incomes were similarly distributed between cases and controls, which suggests that there was successful frequency matching on telephone exchange, and adjustment for education, income, and race did not substantially affect our findings.
The procedures followed were in accordance with each hospitals institutional guidelines. The study was approved by each hospitals institutional review committee, and all participants gave informed consent for participation. For the analyses presented here, only women with complete information on demographic factors and reproductive history were considered. A total of 762 cases and 1,357 controls had complete demographic and reproductive history data. Of these persons, 176 cases and 124 controls were nulligravid (never pregnant). Of the remaining 586 gravid (ever-pregnant) cases and 1,233 gravid controls, four cases and five controls were pregnant for the first time at study enrollment, three cases and three controls reported having had only a tubal or molar pregnancy, and 16 cases and 36 controls reported a history of both spontaneous and induced abortions; these women were excluded from the analyses. Thus, 739 cases and 1,313 community controls were included.
Data collection
A standardized 1.5-hour in-person interview of cases and controls provided detailed information on each participants medical history, general demographic data, family history, and gynecologic and obstetric history, including tubal ligation, family history of ovarian cancer, oral contraceptive use and duration (in months), and pregnancy history. A "life calendar" marked with important events that each participant recalled as occurring during her life was used to enhance memory of distant events. The reference date was calculated as 6 months prior to diagnosis (cases) or interview (controls). The baseline questionnaire asked for detailed information on each pregnancy, including the month and year in which each pregnancy started and ended, the gestational length of each pregnancy (in months), and the outcome of each pregnancy, including type of incomplete pregnancy (spontaneous or induced abortion). Induced abortion was defined as any pregnancy termination described as induced or therapeutic. Spontaneous abortion was defined as any incomplete pregnancy described as a miscarriage or any naturally occurring termination of pregnancy.
Statistical analysis
Comparisons between cases and controls were performed by means of the Wilcoxon rank-sum test for continuous measures and the chi-squared test for discrete measures. Fishers exact test for discrete measures was used when expected cell counts were less than five. Because matching was based on frequencies for only two broad criteria (age within 5-year intervals and three-digit telephone exchange), the matching was not preserved in the analyses (44). The primary hypotheses of interest involved testing associations between incomplete pregnancy history (any incomplete pregnancy, spontaneous abortion, induced abortion) and epithelial ovarian cancer among gravid women. In an attempt to further characterize the independent effect of incomplete pregnancy on ovarian cancer risk, we also evaluated associations between incomplete pregnancy history and epithelial ovarian cancer in nulliparous women and nulligravid women. Odds ratios, 95 percent confidence intervals, and tests of significance were conducted for all variables of interest using unconditional logistic regression. Multivariate logistic regression was used to control for covariates that were shown to differ between cases and controls in univariate analyses, in addition to those that are known to be associated with ovarian cancer risk. These covariates were included in multivariate modeling as follows. Age at diagnosis (cases) or interview (controls), number of births (livebirth or stillbirth), and duration of oral contraceptive use were treated as continuous variables. Race (White, other), history of tubal ligation, and family history of ovarian cancer were treated as dichotomous variables. Educational level (less than high school, high school graduation, more than high school) was treated as a polytomous variable. Additional potential confounders (menopausal status, ever smoking, ever drinking alcohol regularly (weekly or more often for 6 months), duration of breastfeeding (months), history of physician-diagnosed infertility, and self-reported income) were considered but not included, since they did not significantly contribute to or alter the results of the final multivariate models.
To further understand the differences between women with a history of spontaneous abortion and women with a history of induced abortion, we compared demographic and reproductive characteristics between these two mutually exclusive groups separately for cases and controls, using the Wilcoxon rank-sum test for continuous measures and the chi-squared test for discrete measures. We used Fishers exact test for discrete measures when expected cell counts were less than five. To address potential differences in rates of incomplete pregnancy across age or race groups, we conducted two separate subgroup analyses; analyses were restricted to a subgroup of women younger than age 55 years at study enrollment and to a subgroup of White women (sample size was insufficient to limit analyses to other racial groups).
Probability values less than or equal to 0.05 were considered statistically significant. All tests of statistical significance were two-tailed. Analyses were performed using SAS software, release 8.0 (SAS Institute, Inc., Cary, North Carolina).
![]() |
RESULTS |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
|
Tables 2, 3, and 4 show the odds ratios for epithelial ovarian cancer according to history of incomplete pregnancy among gravid women. Compared with controls, cases were no more or less likely to have experienced any type of incomplete pregnancy (adjusted odds ratio (OR) = 0.95, 95 percent confidence interval (CI): 0.76, 1.18) and were no more likely to have a spontaneous or induced abortion (table 2). Both for all incomplete pregnancies and for spontaneous and induced abortions separately, the number of incomplete pregnancies and maternal age at first incomplete pregnancy did not alter ovarian cancer risk. Having a spontaneous abortion before a first birth provided significant protection (adjusted OR = 0.47, 95 percent CI: 0.30, 0.75) (table 3), while a reduced but not significant effect was found for having an induced abortion prior to a first birth (adjusted OR = 0.80, 95 percent CI: 0.44, 1.47) (table 4). No association was found for incomplete pregnancies after a first birth (adjusted OR = 0.96, 95 percent CI: 0.72, 1.28) (table 2), and this result was independent of the type of pregnancy loss.
|
|
|
To eliminate potential confounding by parity, we restricted analyses to women with one pregnancy (table 5). Compared with unigravid women with one incomplete pregnancy, unigravid women with one full-term pregnancy were protected from ovarian cancer (adjusted OR = 0.29, 95 percent CI: 0.15, 0.57); these results were also independent of type of incomplete pregnancy (for induced abortion, adjusted OR = 0.31, 95 percent CI: 0.15, 0.65; for spontaneous abortion, crude OR = 0.17, 95 percent CI: 0.04, 0.63). Cell sizes were too sparse for us to adjust for relevant confounders in the model limited to spontaneous abortion or to model women with two or more pregnancies separately. We then restricted analyses to unigravid women who had never breastfed, in order to eliminate potential confounding by breastfeeding. Compared with the same 36 cases and 24 controls with one incomplete pregnancy, women with one full-term pregnancy who had never breastfed (40 cases and 82 controls) were protected from ovarian cancer after adjustment for age and duration of oral contraceptive use (OR = 0.23, 95 percent CI: 0.11, 0.48; data not shown).
|
No overall change in risk of ovarian cancer related to history of incomplete pregnancy was found when analyses were confined to White women only. Likewise, no overall change in ovarian cancer risk related to gestational length or timing or type of incomplete pregnancy was found when analyses were confined to women aged 55 years or younger at study enrollment (n = 815 controls) or at ovarian cancer diagnosis (n = 318 cases) (data not shown).
![]() |
DISCUSSION |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|
Many studies that have evaluated the association between incomplete pregnancies and the risk of ovarian cancer have reported a significant inverse relation (10, 1924) much like the relation of parity with ovarian cancer (215). Riman et al. (18) conducted an extensive review of the epidemiologic evidence in 1998, concluding that incomplete pregnancies probably reduce ovarian cancer risk, though not to the degree of full-term pregnancies. Although results from the present study do not support an association between number of incomplete pregnancies and ovarian cancer risk, we did observe that an incomplete pregnancy occurring prior to a first birth was protective. One full-term pregnancy was also more protective than one incomplete pregnancy among unigravid women. These results are consistent with etiologic hypotheses of ovarian cancer relating to the suppression of ovulation and pituitary gonadotropin levels during pregnancy (4, 16, 17, 39, 45, 46).
We evaluated spontaneous and induced abortion separately, in order to assess any differential effects of type of incomplete pregnancy on ovarian cancer risk. Our results are consistent with the findings of studies that observed no significant association between ovarian cancer risk and number of incomplete pregnancies, either spontaneous (24, 8, 9, 13, 26, 29, 3234, 3639) or induced (5, 8, 9, 13, 26, 29, 3437, 39). In contrast with other studies that observed a reduced risk associated with induced abortion (6, 21, 38), our results suggest that a gestational length of 3 months prior to a first induced abortion may be associated with increased risk. To our knowledge, only two studies have assessed the relation between gestational length prior to a first induced abortion and ovarian cancer risk, and no association was found (29, 37). Alternatively, Chen et al. (37) observed an 80 percent elevation in risk among women who reported a gestational length of
3 months prior to a first spontaneous abortion (adjusted OR = 1.8, 95 percent CI: 1.1, 3.0). We did not observe a significant relation between gestational length prior to a first spontaneous abortion and ovarian cancer.
Although it may be biologically plausible that spontaneous and induced abortions have different relations with ovarian cancerfor example, through differential effects on inflammation (42, 4754)it is equally likely that this single association may have been a spurious finding due to small subsample sizes and multiple statistical comparisons.
Overall, women in our study who had had an induced abortion appeared to be significantly different from women who had experienced a spontaneous abortion with respect to several ovarian cancer risk factors. In particular, the women with an induced abortion tended to be less parous, younger, more educated, and more racially diverse than women with a spontaneous abortion. Hence, the observed risk associated with increased gestational length prior to a first induced abortion could potentially have been confounded by the demographic differences between the two groups. In multivariate analyses, however, we also adjusted for number of births, age, education, and race; furthermore, no overall change in risk of ovarian cancer related to gestational length prior to a first incomplete pregnancy was found when analyses were restricted to White women only.
Relations between incomplete pregnancy and ovarian cancer risk have been inconsistent, both in previous studies and in the current study. A biologic mechanism that accounts for the well-established protective effects of parity, in addition to allowing for the inconsistent findings regarding incomplete pregnancy, remains to be determined. Nevertheless, such analyses have been helpful in furthering our understanding of the biology of other female cancers. For example, with breast cancer, studies have shown that incomplete pregnancies, regardless of type, are not associated with an increase in breast cancer risk (5562). Further attention to biologic mechanisms might prove useful in advancing our understanding of how interrupted pregnancies influence ovarian cancer risk.
Some limitations of this study deserve consideration. Because of the sensitive nature of incomplete pregnancy, underreporting (recall bias) is one concern. Although strengths of this study included the use of life calendars and structured interviews to enhance recall of reproductive experiences, we relied on self-reports and did not confirm pregnancy outcomes by checking medical records. Spontaneous abortions that occur during the first few weeks of gestation may not be recognized. In addition, women may have incorrectly recalled past events, such as gestational length of their first incomplete pregnancy. If cases and controls were equally likely to underreport their incomplete pregnancy history, the nondifferential misclassification could conceal a true relation of a small magnitude. On the other hand, differential reporting of induced abortions, for instance, could result in a false inflation of the true estimated risk if cases were more likely to report/recall induced abortion history than controls (59). Previous case-control studies of breast and cervical cancers and induced abortion found no evidence of differential reporting of prior induced abortion (55, 63, 64).
The possibility also exists that there are differences in incomplete pregnancy rates across age groups due to differences in the availability of legalized abortion during a womans reproductive years. In the United States, induced abortion was legalized with the 1973 Roe v. Wade Supreme Court decision. Therefore, in a subgroup analysis, we excluded those women who were older than age 55 years at study enrollment or cancer diagnosis (resulting in a maximum age of 34 years in 1973, since study recruitment occurred in 19941998). In spite of the reduced sample size, our data did not suggest any difference in relative odds related to gestational length or timing or type of incomplete pregnancy by age at enrollment or diagnosis (data not shown).
In conclusion, incomplete pregnancy does not appear to be associated with ovarian cancer risk. In addition, our results suggest similar effects of spontaneous and induced abortion on ovarian cancer risk among ever-pregnant women. The possibility of increased risk with increased gestational length in induced abortion requires replication with verified medical records.
![]() |
ACKNOWLEDGMENTS |
---|
![]() |
NOTES |
---|
![]() |
REFERENCES |
---|
![]() ![]() ![]() ![]() ![]() ![]() ![]() |
---|