Abstract
Results of studies assessing intrauterine device (IUD) use and ovarian cancer risk are inconsistent. We examined the association between IUD use, including duration, type, and timing of use, and ovarian cancer risk using three population-based studies. Data from the New England Case-Control Study (NEC) and two prospective cohort studies, the Nurses’ Health Studies (NHS/NHSII), were included in the analysis. Information on IUD use was collected by in-person interview in NEC and by biennial questionnaire in NHS/NHSII. We used unconditional logistic regression to calculate odds ratios (OR) and 95% confidence intervals (CI) in NEC and Cox regression to calculate hazard ratios (HR) and 95% CI in NHS/NHSII. We used meta-analysis to combine the NEC and the pooled NHS/NHSII results. Overall, IUD use was not associated with epithelial ovarian cancer risk (OR=0.96, 95% CI: 0.81–1.14 in NEC; HR=0.89, 95% CI: 0.69–1.15 in NHS/NHSII; combined RR=0.94, 95% CI: 0.81–1.08). Among IUD users, older age at first use was associated with increased ovarian cancer risk (p-trend=0.03). We did not observe significant associations by IUD type or duration of use. In conclusion, IUD use was not associated with ovarian cancer risk in our study.
Keywords: cancer prevention, non-hormonal intrauterine device use, ovarian cancer risk
Introduction
An estimated 60% of American women currently use contraception.1,2 Past studies have suggested that some contraceptive methods strongly influence ovarian cancer risk, particularly oral contraceptives (OC) which decreased ovarian cancer risk by 15–30%.2–4 Moreover, the increase in OC use between the 1970s and 1990s in the United States (U.S.) is likely responsible for the decrease in ovarian cancer incidence over recent decades.2,4,5 While contraception use remains high, there has been a shift in the type of contraception used in the U.S. from OCs to intrauterine device (IUD) since 1990s.1 Furthermore, in 2009 and later reaffirmed in 2018, the American College of Obstetricians and Gynecologists recommended IUDs and contraceptive implants as the most effective and safest forms of reversible contraception for women and adolescents at risk of unwanted pregnancies.6,7
The results of studies on IUD use and ovarian cancer risk are inconsistent. Most studies reported an inverse association between IUD use and ovarian cancer risk,8–12 while others observed increasing risk13 or no association.14–16 The inconsistent results could be due in part to the differences in the IUD types being used between study populations, including a recent shift in trend from non-hormonal to hormonal IUDs.17,18 However, given that earlier studies of IUD use and ovarian cancer risk were predominantly before the introduction of hormonal IUDs, the differences are largely restricted to non-hormonal IUD types. Inconsistent results across studies may also be attributable to differences in duration and timing of IUD use. Moreover, studies have suggested that ovarian cancer is a heterogeneous disease with differences in risk factors by histotype, further leading to variability across studies.19–24
We examined the association between IUD use and ovarian cancer risk in three population-based studies, using data from the New England Case-Control Study (NEC), the Nurses’ Health Study (NHS), and Nurses’ Health Study II (NHSII). Given the timing of the three studies and the age distribution of the study participants included in our analysis, most of the IUD use was non-hormonal. While hormonal IUD is increasing, use of non-hormonal IUDs still remains high in developing countries due to limited access to and higher cost of hormonal IUDs.25,26 Thus, non-hormonal IUD remains an important alternative worldwide for women seeking effective inexpensive long-term reversible contraception without hormones. Understanding the association between IUD use and ovarian cancer risk is critical for establishing the risk associated with this contraceptive approach.27–30
Materials and Methods
New-England Case-Control Study (NEC)
NEC is a population-based case-control study of ovarian cancer. Details of the NEC study design have been described elsewhere.19,31 Briefly, participants were enrolled during four phases: 1984–1988, 1992–1997, 1998–2003, and 2003–2008. Women aged 18–80 years residing in Eastern Massachusetts and New Hampshire who were diagnosed with epithelial ovarian cancer were identified as potential cases through tumor board and registries. Cases were considered ineligible if they had died, moved outside the study area, did not have a telephone number, or had a non-ovarian primary tumor. From the four study phases, 2,465 eligible women with ovarian cancer consented for participation and were subsequently enrolled as cases. Of the 2,465 enrolled cases, we excluded 142 non-epithelial cases from the analysis. Potential controls were identified through random digit dialing (1992–1997 phase only), driver license lists, and town resident lists. Controls were considered ineligible if they had died or moved, were severely ill, did not have a working telephone or speak English, or did not have both ovaries. Across the four study phases, 2,339 eligible women (55% of those eligible through lists and 72% of those eligible through random digit dialing) consented for participation and were enrolled as controls. Controls were frequency-matched to the cases based on age and region of residence. Study participants were interviewed at the time of enrollment. Our final analytic sample in NEC included 2,323 cases and 2,339 controls. This study was approved by the institutional review boards of Brigham and Women’s Hospital, Boston, Massachusetts and Dartmouth College, Hanover, New Hampshire.
Ovarian cancer assessment in NEC
Detailed clinical information on ovarian cancer, including histotype, stage, and grade, was obtained from case pathology reports for all the cases. Pathology reports were then reviewed by a gynecologic pathologist who classified the tumor by grade and histotype (serous borderline, mucinous borderline, low-grade serous invasive, high-grade serous invasive, mucinous invasive, endometrioid, clear-cell, and other).
IUD use and covariate assessment in NEC
During the in-person interview, study participants were queried for information on use of contraception, including IUD use that occurred up to one year prior to the diagnosis for cases and one year prior to the interview for controls. The primary exposure of interest was self-reported ever use of IUD. Among ever IUD users, details on IUD types and duration of use (year/month started; year/month ended) for each IUD type were ascertained. We asked women whether the following type(s) of IUD had ever been used, including copper IUD, Lippes, Dalkon Shield, safety coil, other inert IUDs, unknown IUDs, and progesterone-releasing IUDs (only for phases 1998–2003 and 2003–2008). Duration of IUD use (months) was calculated by summing all the durations reported for each IUD type. Age at first IUD use was calculated as the difference between the earliest year of reported IUD use and year of birth. During the interview, additional information was collected on demographics, health, and reproductive history, including age, race/ethnicity, self-reported body weight and height, age at menarche, number of past pregnancies, OC use (yes if duration of OC use was more than 3 months), type, and duration, family history of ovarian cancer, history of tubal ligation, and menopausal status.
Nurses’ Health Studies (NHS/NHSII)
NHS and NHSII are two large prospective U.S. cohort studies. Details on NHS and NHSII have been described elsewhere.32–34 Briefly, NHS enrolled 121,700 female registered nurses of ages 30 to 55 years in 1976; NHSII enrolled a younger cohort of 116,429 female nurses of ages 25 to 42 years in 1989. All women have been followed up biennially to collect data on lifestyle, health factors, and disease outcomes. For our analysis, we excluded participants with a previous diagnosis of cancer (other than non-melanoma skin cancer), bilateral oophorectomy, or menopause due to pelvic irradiation at baseline. After exclusion, 110,450 NHS participants and 113,104 NHSII participants (total N=223,554) were included in the analysis. Participants were followed from the first questionnaire return date (1976 for NHS and 1989 for NHSII), until the first of the following events: ovarian cancer diagnosis, diagnosis of other cancer (excluding non-melanoma skin cancer), bilateral oophorectomy, pelvic irradiation, death, return of the last questionnaire, or end of the follow-up period (May 31, 2016 for NHS and May 31, 2017 for NHSII; follow-up rates > 90%). We pooled NHS and NHSII cohorts since there was no significant heterogeneity after running a meta-analysis with random effects for cohorts (p-value for cohort heterogeneity: overall: p=0.76). The study protocol was approved by the institutional review boards of the Brigham and Women’s Hospital and Harvard T.H. Chan School of Public Health, and those of participating registries as required.
Ovarian cancer assessment in NHS/NHSII
We identified incident cases of epithelial ovarian cancer either by self-report on the biennial questionnaire or when a participant was reported deceased by a family member or the U.S. Postal Service, with the cause of the death identified via linkage to the National Death Index. We requested permission to review the medical records and pathology reports from participants with a diagnosis of ovarian or primary peritoneal cancer. A gynecologic pathologist reviewed the records to confirm ovarian cancer diagnosis and classified the tumor by grade and histotype. When pathology reports could not be obtained, information on the diagnosis was accessed via linkage to cancer registries.
IUD use and covariate assessment in NHS/NHSII
Among NHS and NHSII participants, demographic information and reproductive and other health-related characteristics were queried at baseline and subsequently every 2 to 4 years through mailed questionnaires. Regular IUD use (yes/no) was queried biennially from 1976 to 1982 among NHS participants and from 1989 to 2009 among NHSII participants. On the questionnaire, the participant was asked “do you currently use any of these forms of contraception?”, with IUD (loops or coils) being listed as one of the options. Duration of IUD use and age at first IUD use were not assessed. Additional reproductive events were regularly updated, including pregnancy, OC use, tubal ligation, menopausal status, and family history of ovarian cancer.
Statistical analysis
IUD use was modeled as ever versus never IUD use in relation to ovarian cancer risk. In NEC, IUD use was coded as ever if a participant reported any IUD use and never otherwise. In NHS/NHS II, IUD use was assessed every two to four years and the exposure information was updated over time. Participants were coded as users when she first reported “yes” for IUD use at a queried cycle and carried forward thereafter, and never otherwise. We first examined the risk of any epithelial ovarian cancer with IUD use. For the NEC study, we used unconditional logistic regression to calculate odds ratio (OR) and 95% confidence interval (CI). For the pooled analysis of NHS/NHSII, we used Cox proportional hazard model stratified on age, time period, and cohort to calculate hazard ratio (HR) and 95% CI for risk of ovarian cancer. We subsequently examined epithelial ovarian cancer risk by the status of invasiveness (borderline vs. invasive) and histotype. In NEC, we used polytomous logistic regression to calculate histotype-specific ORs and 95% CIs. Statistical differences in the histotype-specific associations were evaluated by the likelihood ratio test comparing a model with the association constant across histotypes to a model allowing the association to vary by histotype. In NHS/NHSII, histotype-specific HR and 95% CI were reported only for high-grade serous invasive tumor given the small numbers of the exposed cases for the other histotypes. We adjusted for covariates hypothesized a priori as potential confounders in the analysis, including age (fixed effect in NEC, random effect in NHS/NHSII), body mass index (BMI in kg/m2; <18.5, 18.5–24.9, 25–29.9, ≥30), age at menarche (<12 years, 12 years, >12 years), parity (nulliparous, 1, 2, 3, ≥4), OC use (months), history of tubal ligation (yes/no), family history of ovarian cancer (yes/no), and study center and study phase for NEC. BMI, parity, OC use, and history of tubal ligation were updated at each questionnaire cycle in NHS/NHS II analyses. Finally, we used random effects meta-analysis to calculate the summary effect estimates for the association between IUD use and overall ovarian cancers (borderline and invasive cases combined), borderline ovarian cancers, and invasive ovarian cancers and heterogeneity by study (NEC and NHS/NHSII) and by invasiveness (borderline and invasive).
We subsequently investigated the association between IUD type (copper IUDs, Lippes, Dalkon Shield, safety coil, and other inert/unknown IUDs), age at first IUD use, and duration of IUD use with ovarian cancer risk in the NEC study. Across the four study phases, four progesterone-releasing IUD users were identified (1 from phase 1998–2003 and 3 from phase 2003–2008) and were subsequently grouped into the other inert/unknown IUDs group. Age at first IUD use was categorized into the following groups: <25 years (reference), 25–29 years, 30–34 years, and ≥35 years. Duration of IUD use was categorized into the following groups: never use (reference), ≤12 months, 12–48 months, and >48 months. Multivariable unconditional logistic regression models adjusting for the same set of covariates described above were used to calculate ORs and 95% CIs. Among the IUD users, median values of each age and duration group were entered into the multivariable model to estimate p-values for trend for age at first IUD use and duration of IUD use, respectively.
Additional sensitivity analyses included investigation of potential effect modification by BMI at age 18, parity, OC use, and menopausal status, and potential confounding by history of hysterectomy. Effect modification was evaluated using the likelihood ratio test comparing two models with and without the interaction term between IUD use (yes/no) and the potential modifier of the interest. All analyses were conducted using SAS statistical software (version 9.4; SAS Institute Inc, Cary, NC, USA) and STATA (version 12.1; StataCorp, TX, USA). All statistical tests were 2-sided, with p-values less than 0.05 considered statistically significant.
Results
The NEC analysis included 2,323 epithelial ovarian cancer cases (491 borderline and 1,832 invasive) and 2,339 controls. The pooled NHS/NHSII analysis included 223,554 participants, of which 1,480 incident epithelial ovarian cancer cases (1,003 invasive, 92 borderline, and 34 unknown from NHS; 277 invasive, 58 borderline, and 16 unknown from NHSII) were identified during the follow-up period (5,630,520 person-years).
In NEC, 317 (14%) cases and 376 (16%) controls were identified as ever IUD users, respectively (Table 1, Supplement Table 1). Among the NEC controls who were ever IUD users, the mean age at first IUD use was 27.5 years (SD=5.5 years). With respect to duration of IUD use, 30% reported ≤12 months, 34% reported 12–48 months, and 35% reported >48 months. Among the NEC controls, compared to never IUD users, IUD users were more likely to be parous, to have ever used OCs and had tubal ligation. In the pooled NHS/NHSII cohorts, there were 7,925 women (3.5%) identified as ever IUD users at baseline. Compared to never IUD users, IUD users at baseline were younger, more likely to be parous, more likely to have reported ever OC use, and less likely to have had tubal ligation (Table 1).
Table 1:
Baseline population characteristics of NEC controls (1984–2008) and NHS/NHSII participants (by IUD use1)
| NEC Controls (n=2,339)2 | Pooled NHS and NHS II (n=223,554) | |||
|---|---|---|---|---|
| Never IUD users (n=1,963, 84%) | Ever IUD users (n=376, 16%) | Never IUD users (n=215,629, 97%) | Ever IUD users (n=7,925, 3%) | |
|
| ||||
| Study center, n (%) | ||||
| MA | 1633 (83) | 315 (84) | ||
| NH | 330 (17) | 61 (16) | ||
| Study phase, n (%) | ||||
| 2 | 217 (11) | 22 (6) | ||
| 3 | 451 (23) | 71 (19) | ||
| 4 | 586 (30) | 135 (36) | ||
| 5 | 709 (36) | 148 (39) | ||
| Baseline age (years), mean (SD) | 52.1 (13.2) | 53.1 (9.4) | 38.1 (7.2) | 36.5 (5.0) |
| Baseline body mass index (BMI, kg/m2), means (SD) | 25.9 (5.5) | 25.8 (5.4) | 23.9 (4.6) | 23.4 (4.2) |
| Age at menarche (years), mean (SD) | 12.7 (1.6) | 12.7 (1.4) | 12.4 (1.6) | 12.4 (1.6) |
| Parity, n (%) | ||||
| Nulliparous | 387 (20) | 34 (9) | 38,715 (18) | 415 (5) |
| 1 | 248 (13) | 46 (12) | 29,259 (14) | 889 (11) |
| 2 | 570 (29) | 156 (41) | 65,309 (30) | 3,282 (41) |
| 3 | 384 (20) | 77 (20) | 46,128 (21) | 2,044 (26) |
| 4 | 374 (19) | 63 (17) | 36,218 (17) | 1,295 (16) |
| Oral contraceptive use, n (%) | ||||
| Never use | 823 (42) | 86 (23) | 75,907 (35) | 2,331 (29) |
| <2 years | 277 (14) | 95 (25) | 41,788 (19) | 2,214 (28) |
| 2–5 years | 298 (15) | 94 (25) | 45,333 (21) | 2,081 (26) |
| ≥5 years | 565 (29) | 101 (27) | 52,601 (24) | 1,299 (16) |
| Tubal ligation, n (%) | 316 (16) | 117 (31) | 29,310 (14) | 26 (0.3) |
| Premenopausal, n (%) | 885 (45) | 159 (42) | 193,510 (90) | 7,872 (99) |
| Family history of ovarian cancer, n (%) | 44 (2) | 11 (3) | 4,051 (2) | 163 (2) |
| Smoking status, n (%) | ||||
| Never | 953 (49) | 155 (41) | 118,713 (55) | 3,688 (47) |
| Former | 708 (36) | 165 (44) | 47,748 (22) | 1,883 (24) |
| Current | 302 (15) | 56 (15) | 49,168 (23) | 2,354 (30) |
Time of enrollment for the four NEC study phases ranges from 1988 to 2008; NEC cases were enrolled at the time of ovarian cancer diagnosis, and controls were matched to the cases on age and region of residence. For NHS and NHSII studies, year of baseline was 1976 for NHS and 1989 for NHSII, respectively. Given the timing of the studies, among the reported IUD users, the majority of them were non-hormonal IUD users.
Characteristics of NEC cases by IUD use were summarized in Supplement Table 1.
Multivariable-adjusted results from the NEC study, NHS/NHSII studies, and the meta-analysis of the three studies combined showed little evidence of an association between IUD use and risk of epithelial ovarian cancer overall with no significant heterogeneity by study (NEC: OR=0.96, 95% CI: 0.81–1.14; NHS/NHSII: HR=0.89, 95% CI: 0.69–1.15; pooled RR=0.94, 95% CI: 0.81–1.08; p-heterogeneity by study=0.65; Table 2, Figure 1).
Table 2:
Ever use of IUD and ovarian cancer risk, overall and by tumor invasiveness and histotype
| NEC | NHS & NHSII | ||||||
|---|---|---|---|---|---|---|---|
| Never IUD use | Ever IUD use | OR (95% CI) | Never IUD use | Ever IUD use | HR (95% CI) | ||
|
| |||||||
| Overall | Person-years | N/A | N/A | 5,349,694 | 280,826 | ||
| Number of cases | 2,006 | 317 | 1,416 | 64 | |||
| Age-adjusted | 0.83 (0.70–0.97) | 0.90 (0.70–1.16) | |||||
| Multivariate1,2 | 0.96 (0.81–1.14) | 0.89 (0.69–1.15) | |||||
|
| |||||||
| Borderline | Person-years | N/A | N/A | 5,348,318 | 280,767 | ||
| Number of cases | 414 | 77 | 140 | 10 | |||
| Age-adjusted | 1.12 (0.85–1.47) | 1.27 (0.66–2.44) | |||||
| Multivariate1,2 | 1.30 (0.98–1.73) | 1.23 (0.64–2.38) | |||||
|
| |||||||
| Invasive | Person-years | N/A | N/A | 5,349,485 | 280,815 | ||
| Number of cases | 1,592 | 240 | 1,228 | 52 | |||
| Age-adjusted | 0.79 (0.66–0.94) | 0.85 (0.64–1.13) | |||||
| Multivariate1,2 | 0.90 (0.75–1.09) | 0.85 (0.64–1.12) | |||||
|
| |||||||
| Borderline1,3 | Serous borderline | 241 | 49 | 1.43 (1.01–2.02) | |||
| Mucinous borderline | 150 | 19 | 0.90 (0.54–1.50) | ||||
| Pheterogeneity | 0.11 | ||||||
|
| |||||||
| Invasive1,3 | Low-grade serous invasive | 53 | 6 | 0.65 (0.28–1.54) | |||
| High-grade serous invasive | 901 | 150 | 0.95 (0.77–1.18) | 0.92 (0.67–1.27) | |||
| Mucinous invasive | 102 | 16 | 1.11 (0.64–1.93) | ||||
| Endometrioid | 318 | 44 | 0.91 (0.65–1.29) | ||||
| Clear-cell | 122 | 9 | 0.55 (0.27–1.10) | ||||
| Pheterogeneity | 0.37 | ||||||
Multivariable model for NEC analysis was adjusted for age (years), study center, study phase, body mass index in kg/m2 (<18.5, 18.5–24.9, 25–29.9, ≥30), age at menarche (<12 years, 12 years, >12 years), parity (nulliparous, 1, 2, 3, ≥4), oral contraceptive use (months), tubal ligation, and family history of ovarian cancer.
Multivariable model for pooled NHS/NHS II analysis was stratified on age, time period, and cohort, and adjusted for body mass index in kg/m2 (<18.5, 18.5–24.9, 25–29.9, ≥30), age at menarche (<12 years, 12 years, >12 years), parity (nulliparous, 1, 2, 3, ≥4), oral contraceptive use (months), tubal ligation, and family history of ovarian cancer.
Histotype-specific ORs and 95% CIs from multivariable polytomous logistic regression adjusting for the same set of covariates were calculated within borderline tumors and invasive tumors, separately.
Figure 1: Meta-analysis of NEC and NHS/NHSII results.
Summary risk ratios and 95% confidence intervals for the association between intrauterine device use and ovarian cancer risk of overall, borderline, and invasive tumors of NEC and NHS/NHSII studies.
When examining IUD use and risk of ovarian cancer by the status of invasiveness, we observed a non-significant increase in risk of borderline ovarian tumors (pooled RR=1.29, 95% CI: 0.99–1.67) and a non-significant decrease in risk of invasive ovarian tumors (pooled RR=0.89, 95% CI: 0.76–1.03), and there was no significant heterogeneity by invasiveness (p-value for heterogeneity: p=0.12; Figure 1). Directions of the associations were consistent in the separate analyses on NEC and NHS/NHSII studies for both borderline tumors (NEC: OR=1.30, 95% CI: 0.98–1.73; NHS/NHSII: HR=1.23, 95% CI: 0.64–2.38) and invasive tumors (NEC: OR=0.90, 95% CI: 0.75–1.09; NHS/NHS II: HR=0.85, 95% CI: 0.64–1.12; Table 2). We further examined the association by histology within borderline and invasive tumors in NEC. While the associations were not significantly different across histology within the borderline tumors and the invasive tumors (p-value for heterogeneity: p=0.11 for borderline tumors; p=0.37 for invasive tumors), IUD use was associated with an increased risk of serous borderline tumors in NEC (OR=1.43, 95% CI: 1.01–2.02; Table 2). Due to the low prevalence of exposure (IUD users) and limited numbers of histotype-specific ovarian cancer cases, we were only able to examine the histotype-specific association for high-grade serous invasive tumors in NHS/NHSII; estimated risks for high-grade serous invasive tumors were similar between the NEC and NHS/NHSII studies (NEC: OR=0.95, 95% CI: 0.77–1.18; NHS/NHSII: HR=0.92, 95% CI: 0.67–1.27;Table 2).
When examining detailed patterns of IUD use in NEC, no association was observed between any of the IUD types and ovarian cancer risk (Table 3). Similarly, no evidence of association was observed between duration of IUD use and risk of ovarian cancer. However, among ever IUD users, there was a statistically significant trend of increased ovarian cancer risk with older age at first IUD use (compared to women with age at first IUD use <25 years, OR=0.98 for 25–29 years,1.19 for 30–34 years, and 1.81 for ≥35 years; p-trend = 0.03; Table 3).
Table 3:
Characteristics of IUD use and ovarian cancer risk in NEC
| n (cases/controls) | OR1 (95% CI) | |
|---|---|---|
|
| ||
| IUD type | ||
| Never IUD use | 2006/1963 | ref |
| Copper IUDs2 | 105/110 | 1.04 (0.78–1.38) |
| Lippes | 25/38 | 0.72 (0.43–1.22) |
| Dalkon Shield | 37/45 | 0.96 (0.61–1.51) |
| Safety coil | 11/14 | 0.95 (0.42–2.13) |
| Other inert/unknown IUDs3 | 139/169 | 0.96 (0.76–1.22) |
| Age at first IUD use 4 | ||
| <25 | 114/117 | ref |
| 25–29 | 100/132 | 0.98 (0.66–1.46) |
| 30–34 | 64/83 | 1.19 (0.73–1.94) |
| ≥35 | 36/40 | 1.81 (0.95–3.44) |
| p-trend | 0.03 | |
| Duration of IUD use 5 | ||
| Never IUD use | 2006/1963 | ref |
| ≤12 months | 97/113 | 1.00 (0.75–1.34) |
| 12–48 months | 88/128 | 0.81 (0.61–1.08) |
| >48 months | 130/132 | 1.06 (0.82–1.37) |
| p-trend6 | 0.73 | |
Multivariable model adjusted for age (years), study center, study phase, body mass index in kg/m2 (<18.5, 18.5–24.9, 25–29.9, ≥30), age at menarche (<12 years, 12 years, >12 years), parity (nulliparous, 1, 2, 3, ≥4), oral contraceptive use (months), tubal ligation, and family history of ovarian cancer.
Copper IUD included Copper 7, Copper T, and unknown Copper IUDs.
Other inert/unknown IUDs included 4 progesterone-releasing IUD users.
3 cases and 4 controls were excluded from the analysis due to missing value of age at first IUD use.
2 cases and 3 controls were excluded from the analysis due to missing value of duration of IUD use.
P-trend for duration of IUD use was among IUD users.
In sensitivity analyses examining potential effect modification by important covariates, the association between IUD use and ovarian cancer risk generally did not differ by any of the covariates that we examined, including BMI at age 18, parity, OC use, or menopausal status (Supplement Table 2). Further adjusting for hysterectomy in the main association of interest did not influence the results (results not shown).
Discussion
In this study, we evaluated history and patterns of IUD use, including IUD type, timing, and duration of use, in relation to ovarian cancer risk overall and by tumor invasiveness and histotype. Overall, we did not observe significant associations, although there was a suggestion of a possible increased risk for women who initiated IUD use at a later age. The ability to adjust for detailed reproductive and contraceptive history and the consistent results from NEC and the pooled NHS/NHSII analyses further add strength to our study over prior study reports. Given the timing of the three studies included in our analysis, our findings can only be generalized to non-hormonal IUD types.
We previously examined IUD use and risk of invasive ovarian cancer using data from three NEC phases (1992–2008) and found similar null findings for invasive ovarian tumors (RR=0.85, 95% CI: 0.60–1.20 for rapidly fatal invasive tumors with death within 3 years of diagnosis; RR= 0.96, 95% CI: 0.78–1.18 for all other invasive tumors).16 In NHS, we previously examined IUD use and ovarian cancer risk with 612 incident ovarian cancer cases accrued during 28 years of follow-up (1976–2004) and reported an increased risk of ovarian cancer with IUD use (HR=1.76, 95% CI: 1.08–2.85).13 Compared to our earlier studies, the results presented here included additional 183 cases from NEC during phase 1984–1988 and 868 ovarian cancer cases and a longer follow-up period up to 40 years in NHS/NHSII, leading to greater statistical power and allowing us to examine longer latency periods.
Most of the earlier studies of IUD and ovarian cancer risk reported inverse9–11 or null association,14,15 except the earlier NHS study which reported a positive association.13 While results from the prospective Shanghai Women’s Health Study (SWHS) showed no association,14,15 there were several case-control studies9–11 conducted in the similar time range reporting an inverse association between IUD use and risk of ovarian cancer. Ness and colleagues conducted two U.S-based case-control studies (727 ovarian cancer cases in the study from 1994–1999 and 902 ovarian cancer cases in the study from 2003–2008) in which they observed about a 20% decreased risk of ovarian cancer for IUD users.10,11 Another case-control study conducted in Vietnam between 2001–2006 also reported reduced risk of ovarian cancer with IUD use (OR=0.45, 95% CI: 0.33–0.60).9
With respect to IUD use characteristics, we did not see a difference in risk of ovarian cancer by IUD type or duration use. Two studies have evaluated detailed patterns of IUD use. The study conducted by Ness et al. reported a decreased risk with shorter duration of IUD use (IUD use ≤4 years vs. never use: OR=0.53),10 whereas the SWHS study by Huang et al. reported a decreased risk with longer duration of IUD use (IUD use ≥20 years vs. never use: HR=0.62).15 Given the mixed findings on duration of IUD use, we hypothesize that the differences observed across different studies may be due to different contraceptive use pattern and different distribution of ovarian cancer histotypes. Furthermore, types of IUD use and ovarian cancer histotypes are likely to differ, particularly when comparing these characteristics between the U.S. populations and Asian populations. Interestingly, we observed a trend in increasing ovarian cancer risk with older age at first use, but power to evaluate the association at older ages was limited. In the SWHS study, investigators did not observe differences in risk of ovarian cancer by age at IUD initiation <30 or ≥30 compared to never IUD users.14
Regarding the increased risk of ovarian cancer with older age at first IUD observed in this study, while the biological mechanisms behind this observation are unclear, several past studies have suggested the timing of reproductive events, such as menarche35, oral contraceptive use36, pregnancy36–40, and lactation41, 42, might be associated with risk of ovarian cancer. Explanations for why risk factors for ovarian cancer vary with age include changes in cellular state and function43, 44, immune function45, 46, inflammation47, 48, and metabolic function49 with age. Thus, our observed trend in increased risk of ovarian cancer with older age at first IUD use in NEC suggest that IUD use may have differential impact on ovarian cancer risk by age at when women are exposed.
In the stratified analyses, we observed statistical difference by the status of OC use in the NHS/NHS II studies, particularly an inverse association among never OC users. However, given the limited sample size, the potential effect modification by OC use needs to be interpreted with caution.
Our study has several strengths. First, our study included two types of epidemiological studies with distinct strengths. NEC, a population-based case-control study, included detailed information on type, duration, and timing of IUD use and a large enough sample size to enable histotype-specific analyses. NHS/NHSII, which are prospective cohort studies, allowed us to examine incident ovarian cancer cases by following people for long period of time prospectively, avoiding the possibility of recall bias. Detailed information on important covariates was available in all the three studies, allowing us to sufficiently adjust for potential confounders. Our study has some limitations. First, due to the limited prevalence of the exposure, we were not able to assess the association by histotype in NHS/NHSII. Secondly, with respect to exposure misclassification, similar to other case-control studies, NEC was susceptible to recall bias since IUD use was reported after ovarian cancer diagnosis. However, results were similar in NHS/NHSII, and any exposure misclassification in the cohorts would be non-differential with respect to the outcome and thus likely bias the results towards the null. Thirdly, our results can only be generalized to predominantly white populations with similar characteristics of our study sample. Finally, information on the type of IUDs (hormonal vs non-hormonal) was not assessed in the earlier NEC phases (1984–1988 and 1992–1997) or NHS/NHSII studies. Therefore, similar to the earlier studies, our study included a mixture of non-hormonal and hormonal IUDs. Given the timing of the three studies and the age distribution of the participants at baseline, we consider the majority of IUDs were non-hormonal, and thus our results on the association should be interpreted mainly for non-hormonal IUDs. Jareid and colleagues recently examined levonorgestrel-releasing IUD use in a registry-based cohort, and they reported an inverse association between levonorgestrel-releasing IUD use and risk of ovarian cancer (RR=0.53, 95% CI: 0.32–0.88).8 Additional studies with later birth cohorts are needed to fully examine the role of hormonal IUD types with risk of ovarian cancer.
Overall, based on our analysis in NEC and the pooled NHS/NHSII studies, we did not observe evidence of an association between IUD use and ovarian cancer risk. Larger studies are needed to evaluate whether the association differs by histotype, and new studies are needed to evaluate more recent IUD types, particularly hormonal IUDs. Finally, the association between IUD use and ovarian cancer risk needs to be evaluated in diverse study populations to determine whether our findings are generalizable.
Supplementary Material
Novelty and impact:
This study examined IUD use, including duration, type, and timing of the first use, and risk of ovarian cancer by invasiveness and histotype, using three U.S. population-based studies. Overall, we did not find significant association between intrauterine device use and risk of ovarian cancer. Additional investigations are needed to confirm our findings in diverse populations and to evaluate more recent hormonal IUDs with risk of ovarian cancer.
Acknowledgements:
We would like to acknowledge the Channing Division of Network Medicine, Department of Medicine, Brigham and Women’s Hospital and Harvard Medical School, Boston, MA, USA, as the home of the Nurses’ Health Study. The Nurses’ Health Studies would like to thank the following state cancer registries for their help: AL, AZ, AR, CA, CO, CT, DE, FL, GA, ID, IL, IN, IA, KY, LA, ME, MD, MA, MI, NE, NH, NJ, NY, NC, ND, OH, OK, OR, PA, RI, SC, TN, TX, VA, WA, WY. The authors assume full responsibility for analyses and interpretation of these data.
Funding Support:
NIH R01-CA54419, P50-CA105009 and Department of Defense W81XWH-10-1-02802. The Nurses’ Health Studies are supported by NIH grants UM1 CA186107, P01 CA87969, and U01 CA176726. The content is solely the responsibility of the authors and does not necessarily represent the official views of the National Institutes of Health.
Abbreviations:
- BMI
Body mass index
- CI
Confidence interval
- HR
Hazard ratio
- IUD
Intrauterine device
- NEC
New England Case-Control Study
- NHS
Nurses’ Health Study
- NHSII
Nurses’ Health Study II
- OC
Oral contraceptives
- OR
Odds ratio
- RR
Risk ratio
- The U.S.
The United States
Footnotes
Conflict of interest:
Dr. Shelley Tworoger reported receiving grants from NIH/NCI during the conduct of the study. Other authors declared no conflict of interest.
Ethical approval:
NEC study was ethically approved by the institutional review boards of Brigham and Women’s Hospital, Boston, Massachusetts and Dartmouth College, Hanover, New Hampshire. NHS and NHS II studies were ethically approved by the institutional review boards of the Brigham and Women’s Hospital and Harvard T.H. Chan School of Public Health, and those of participating registries as required. Written informed consent was obtained from all study participants in NEC study. Completion of the questionnaire was considered to imply informed consent for NHS and NHS II studies.
Data availability statement:
Further information including the procedures to obtain and access data from the Nurses’ Health Studies is described at https://www.nurseshealthstudy.org/researchers (contact email: nhsaccess@channing.harvard.edu).
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Associated Data
This section collects any data citations, data availability statements, or supplementary materials included in this article.
Supplementary Materials
Data Availability Statement
Further information including the procedures to obtain and access data from the Nurses’ Health Studies is described at https://www.nurseshealthstudy.org/researchers (contact email: nhsaccess@channing.harvard.edu).