Over 22,000 cases of ovarian cancer were diagnosed in 2007 in the United States, but only a fraction of them can be attributed to mutations in highly penetrant genes such as BRCA1. To determine whether low-penetrance genetic variants contribute to ovarian cancer risk, we genotyped 1,536 single nucleotide polymorphisms (SNP) in several candidate gene pathways in 848 epithelial ovarian cancer cases and 798 controls in the North Carolina Ovarian Cancer Study (NCO) using a customized Illumina array. The inflammation gene interleukin-18 (IL18) showed the strongest evidence for association with epithelial ovarian cancer in a gene-by-gene analysis (P = 0.002) with a <25% chance of being a false-positive finding (q value = 0.240). Using a multivariate model search algorithm over 11 IL18 tagging SNPs, we found that the association was best modeled by rs1834481. Further, this SNP uniquely tagged a significantly associated IL18 haplotype and there was an increased risk of epithelial ovarian cancer per rs1834481 allele (odds ratio, 1.24; 95% confidence interval, 1.06-1.45). In a replication stage, 12 independent studies from the Ovarian Cancer Association Consortium (OCAC) genotyped rs1834481 in an additional 5,877 cases and 7,791 controls. The fixed effects estimate per rs1834481 allele was null (odds ratio, 0.99; 95% confidence interval, 0.94-1.05) when data from the 12 OCAC studies were combined. The effect estimate remained unchanged with the addition of the initial North Carolina Ovarian Cancer Study data. This analysis shows the importance of consortia, like the OCAC, in either confirming or refuting the validity of putative findings in studies with smaller sample sizes. (Cancer Epidemiol Biomarkers Prev 2008;17(12):3567–72)

Ovarian cancer is the leading cause of death among cancers of the female reproductive tract and is the fifth leading cause of cancer death in women in the United States (1). The most commonly reported factors that increase ovarian cancer risk are germ-line mutations in the BRCA1 or BRCA2 genes, family history of ovarian cancer, and endometriosis and characteristics such as increasing parity, oral contraceptive use, and tubal ligation reduce risk (2-4). Although the mechanisms underlying these reproductive risk factors have yet to be fully elucidated, it has been hypothesized that they may relate to inflammation and DNA damage (5-7). Ovulation involves disruption of the ovarian surface and is associated with an inflammatory response that involves prostaglandins (8), cytokines (9), and reactive oxygen species (10, 11), all of which have been implicated in carcinogenesis (11-14). The inflammation hypothesis is further supported by the observations that endometriosis may cause inflammation, whereas tubal ligation may reduce inflammation by blocking external agents from coming in contact with the ovary (5). Repair of the ovary after ovulation involves cellular replication that may be prone to DNA damage; thus, errors in DNA repair could also contribute to ovarian cancer. In view of the potential importance of inflammation and DNA repair pathways in ovarian carcinogenesis, inherited variation in genes in these pathways could affect ovarian cancer susceptibility.

To address this hypothesis, we used a candidate gene approach to examine the associations between genes involved in inflammation and DNA repair in a population-based, case-control study of ovarian cancer. When examining numerous single nucleotide polymorphisms (SNP) in multiple genes, the risk of identifying false-positives is present even when statistically adjusting for multiple comparisons. Replication of significant findings using additional independent studies is critical to establish whether true associations exist. In this article, we describe our evaluation of candidate genes related to DNA repair and inflammation followed by replication of putative significant findings within a large international consortium of ovarian cancer case-control studies.

Hypothesis Generating Study

The North Carolina Ovarian Cancer Study (NCO) is a population-based case-control study of epithelial ovarian cancer (borderline and invasive) among women in 48 North Carolina counties. Eligible cases, ages 20 to 74 years, were diagnosed between January 1, 1999 and March 31, 2007 and were identified via rapid case ascertainment through the North Carolina Central Cancer Registry. Population-based controls were identified using list-assisted random-digit dialing and were frequency-matched to the cases by age and race. DNA was obtained from >98% of participants. Further details of the study have been described elsewhere (15). All participants signed informed consent forms and approval for the study was obtained from the Duke University Medical Center Institutional Review Board, participating hospitals, and the Human Subjects Committee at the North Carolina Central Cancer Registry.

Replication Studies

The Ovarian Cancer Association Consortium (OCAC) was formed to provide a forum for researchers to evaluate genetic associations with ovarian cancer with increased power. A major aim of the OCAC is to follow up on promising genetic associations while addressing the problem of multiple comparisons and false discoveries that are inherent to studies using high-throughput genotyping technologies. Twelve OCAC studies from the United States (DOV, HAW, HOP, MAY, STA, UCI, and USC), Europe (GER, MAL, SEA, and UKO), and Australia (AUS) contributed data to the current analysis for a total of 5,877 cases and 7,791 controls of self-reported Caucasian ancestry. These studies have been described in detail previously and are summarized in Table 1 (6, 16-26). All studies obtained approval from their respective human subjects ethics committees and all participants provided signed informed consent.

Table 1.

Description of the NCO and 12 OCAC replication studies with IL18 rs1834481 genotype frequencies (G>C) by case status (listed alphabetically)

StudyLocationCase ascertainmentCases (n)
Control ascertainment*Controls (n)
GGGCCCGGGCCC
AUS (16) Multiregional, Australia Cancer registries, surgical treatment centers 475 359 85 P: Electoral roll 573 410 71 
DOV (17) Washington, USA Cancer Surveillance System, SEER 411 272 55 P: RDD 422 271 46 
GER (18) Multiregional, Germany Hospital admissions 143 89 13 P: Population registries 172 81 13 
HAW (19) Hawaii, USA Hawaii Tumor Registry, SEER 56 30 P: Annual Hawaii Health Survey 91 57 
HOP (20) Ohio, Pennsylvania, and New York, USA Registries, physician offices, pathology databases 174 124 20 P: RDD 367 233 36 
MAL (21) Multiregional, Denmark Danish Cancer Registry, 16 gynecologic departments 232 154 29 P: Danish Central Population Register 643 442 89 
MAY (22) Multiregional, USA Mayo Clinic 221 142 13 C: Women seeking general examinations at Mayo Clinic 245 186 22 
NCO (15) North Carolina, USA North Carolina Central Cancer Registry 450 334 64 P: RDD 470 285 43 
SEA (6) Cambridge, UK East Anglia and West Midlands cancer registries 625 395 71 P: European Prospective Investigation into Cancer and Nutrition-Norfolk cohort 672 456 82 
STA (23) California, USA Greater Bay Area Cancer Registry, SEER 201 102 16 P: RDD 244 141 28 
UCI (24) California, USA Cancer Surveillance Program of Orange County, Tumor Registry 268 159 24 P: RDD 264 172 29 
UKO (25) Multiregional, UK 10 Gynecologic oncology National Health Service Centers 148 291 17 P: UK Collaborative Trial of Ovarian Cancer Screening and all women followed up for cancers by Office of National Statistics 302 231 43 
USC (26) California, USA Los Angeles Cancer Surveillance Program 413 211 37 P: Neighborhood recruits 408 205 36 
Totals — — 3,817 2,462 446 — 4,873 3,170 546 
StudyLocationCase ascertainmentCases (n)
Control ascertainment*Controls (n)
GGGCCCGGGCCC
AUS (16) Multiregional, Australia Cancer registries, surgical treatment centers 475 359 85 P: Electoral roll 573 410 71 
DOV (17) Washington, USA Cancer Surveillance System, SEER 411 272 55 P: RDD 422 271 46 
GER (18) Multiregional, Germany Hospital admissions 143 89 13 P: Population registries 172 81 13 
HAW (19) Hawaii, USA Hawaii Tumor Registry, SEER 56 30 P: Annual Hawaii Health Survey 91 57 
HOP (20) Ohio, Pennsylvania, and New York, USA Registries, physician offices, pathology databases 174 124 20 P: RDD 367 233 36 
MAL (21) Multiregional, Denmark Danish Cancer Registry, 16 gynecologic departments 232 154 29 P: Danish Central Population Register 643 442 89 
MAY (22) Multiregional, USA Mayo Clinic 221 142 13 C: Women seeking general examinations at Mayo Clinic 245 186 22 
NCO (15) North Carolina, USA North Carolina Central Cancer Registry 450 334 64 P: RDD 470 285 43 
SEA (6) Cambridge, UK East Anglia and West Midlands cancer registries 625 395 71 P: European Prospective Investigation into Cancer and Nutrition-Norfolk cohort 672 456 82 
STA (23) California, USA Greater Bay Area Cancer Registry, SEER 201 102 16 P: RDD 244 141 28 
UCI (24) California, USA Cancer Surveillance Program of Orange County, Tumor Registry 268 159 24 P: RDD 264 172 29 
UKO (25) Multiregional, UK 10 Gynecologic oncology National Health Service Centers 148 291 17 P: UK Collaborative Trial of Ovarian Cancer Screening and all women followed up for cancers by Office of National Statistics 302 231 43 
USC (26) California, USA Los Angeles Cancer Surveillance Program 413 211 37 P: Neighborhood recruits 408 205 36 
Totals — — 3,817 2,462 446 — 4,873 3,170 546 

Abbreviations: SEER, Surveillance Epidemiology and End Results; RDD, random-digit dialing.

*

P, population-based; C, clinic-based.

SNP Selection, Genotyping, and Quality Control

The NCO genotyped a total of 1,536 SNPs using the Illumina Golden Gate Assay (Supplementary Table S1). SNPs tagging 120 candidate genes and nonsynonymous SNPs from an additional 50 candidate genes were included on the Illumina OPA. Genes were chosen based on previous literature and were defined at 10 kb upstream and downstream of the gene. Although the DNA repair, inflammation, and hormone candidate gene pathways were predominantly represented, a limited number of genes from the cell cycle, metabolism, methylation, and signal transduction pathways were also included on the Illumina OPA. Tagging SNPs were selected using HapMap version 2028

and ldSelect (27); a minor allele frequency > 0.05 and pairwise linkage disequilibrium threshold of r2 > 0.8 were used for selection. Cases and controls were randomly mixed within each plate and six CEPH-Utah trios-standard by Coriell were distributed across six plates. Only five samples (<1%) failed genotyping. There was one within-plate and one across-plate duplicate sample on each 96-well DNA plate; the concordance rate for these samples was 99.5%.

Eleven of the 12 OCAC studies used the 5′ nuclease TaqMan allelic discrimination assay (TaqMan; Applied Biosystems) to genotype interleukin-18 (IL18) rs1834481 in seven laboratories using a common batch of reagents. One study (AUS) used the iPLEX Sequenom MassArray system (Sequenom). To ensure consistency of genotype calls across laboratories, each site also genotyped a common set of samples from Coriell29

consisting of 90 unique DNA samples (30 CEPH-Utah trios), 5 duplicate samples, and a negative template control. Genotype concordance on these plates was >98%.

Statistical Analyses: NCO

We restricted our analysis to White, non-Hispanic NCO participants for whom genotyping data were available. We performed tests for Hardy-Weinberg equilibrium among the controls for all SNPs using χ2 goodness-of-fit tests and carried out a two-stage analysis. First, for each gene, we fitted an unconditional logistic regression model for disease given age and indicator variables for both the dominant and the recessive genotypes of each SNP in the gene. We used a likelihood ratio test of this model against the model containing only age to assess the degree of association of the gene. We prioritized genes according to the P values of these tests and calculated a q value for each gene to provide estimates of false discovery rates (28). Second, within the top genes, we carried out a Bayesian model selection analysis of the SNP genotype variables using the bic.glm algorithm in the BMA package (29) for the R statistical language to determine the combination(s) giving the best fit. The two-stage analysis was completed using the statistical software package R (30). A haplotype analysis for the gene with the strongest evidence for an association with epithelial ovarian cancer was conducted using Haploview version 3.32 (31).

We completed a full analysis of known and suspected risk factors for epithelial ovarian cancer to determine the extent of confounding; covariates that changed the estimate of effect by ≥10% were considered to be confounders (there were none).

Statistical Analyses: OCAC

IL18 rs1834481 was evaluated by 12 OCAC studies. Among controls, Hardy-Weinberg equilibrium was tested and minor allele frequencies were compared to ensure that there were no important population differences with respect to the allele frequencies of the SNPs. For each OCAC study, we fitted unconditional logistic regression models, adjusting for age, to estimate odds ratios (OR) and 95% confidence intervals (95% CI). Using the same approach to confounding described above, we evaluated the final models for confounding (there was none).

Fixed- and random-effects models accounting for study site were fitted using inverse variance weighting and the DerSimonian-Laird (32) methods, respectively, to estimate summary ORs and 95% CIs for the association between rs1834481 and epithelial ovarian cancer risk. We used Cochran's Q statistic (33) and I2 (34) to evaluate heterogeneity of results. Data were analyzed with and without the NCO data and were restricted to subgroups of cases (all, invasive only, and serous invasive only) because there may be etiologic heterogeneity of epithelial ovarian cancer by histologic type.

The unconditional logistic regression models for the OCAC studies were fitted using SAS (version 9.1.3) and the fixed- and random-effects analyses were fitted using STATA (version 9; StataCorp).

Hypothesis Generating Study: NCO

The initial NCO analysis included 848 epithelial ovarian cancer cases, with a mean age at diagnosis of 55 years, and 798 controls; all participants were White, non-Hispanic. The majority of cases (79%) were invasive and >65% of invasive cases were International Federation of Gynecology and Obstetrics stage III or IV. Serous histology was the most common subtype (61%).

The gene-by-gene analysis identified IL18 as the gene with the strongest evidence for association (that is, smallest P value) with epithelial ovarian cancer (P = 0.002; q value = 0.240) compared with all other genes. Eleven SNPs tagging IL18 had been genotyped in the NCO: rs243908, rs1293344, rs1834481, rs1946519, rs2043055, rs549908, rs5744247, rs5744258, rs5744280, rs4937113, and rs11214109. The Bayesian model selection analysis of the IL18 SNPs identified models that included only rs1834481 genotypes as being the most likely, a posteriori. In a subsequent haplotype analysis of IL18 (Supplementary Table S2), four haplotypes had estimated frequencies of >5%. Only one was significantly associated with epithelial ovarian cancer (P = 0.0007) and was uniquely tagged by rs1834481, which was in Hardy-Weinberg equilibrium (P = 0.981) and had a minor allele frequency of 0.23 in NCO controls.

Using Akaike's Information Criterion, we determined that a log-additive genetic model for rs1834481 best fit the data. There was an increased risk of epithelial ovarian cancer (per allele) for all cases (OR, 1.24; 95% CI, 1.06, 1.45), invasive cases (OR, 1.25; 95% CI, 1.06, 1.28), and serous invasive cases (OR, 1.31; 95% CI, 1.08, 1.60).

Replication Studies: OCAC

Twelve OCAC studies genotyped IL18 rs1834481 in an additional 5,877 cases and 7,791 controls. The SNP was in Hardy-Weinberg equilibrium in all studies (α = 0.10). The minor allele frequencies among controls ranged from 0.20 to 0.28 in these studies with no statistically significant differences between them.

For all of the OCAC studies, the age-adjusted ORs and 95% CIs were not statistically different from a null association (Table 2). In fitting fixed- and random-effects models to the 12 OCAC studies, there was no significant heterogeneity; thus, we report fixed effects here. The fixed-effects estimates did not change when the NCO data were included; there was, however, significant heterogeneity. In a meta-regression, we did not find evidence that the following study characteristics explained this heterogeneity: U.S. versus non-U.S. populations (P = 0.87), incident versus prevalent cases (P = 0.38), and population-based versus clinic-based controls (P = 0.16). Overall, the fixed-effects estimates indicate a null association between IL18 rs1834481 and epithelial ovarian cancer (Table 2; Fig. 1).

Table 2.

Site-specific and fixed-effects summary ORs (per allele), 95% CIs, and heterogeneity statistics for IL18 rs183441 among 12 OCAC studies, excluding NCO

SiteAll cases,* OR (95% CI)Invasive cases,* OR (95% CI)Serous invasive cases,* OR (95% CI)
AUS 1.14 (0.99-1.31) 1.08 (0.93-1.26) 1.17 (0.98-1.39) 
DOV 1.07 (0.91-1.26) 1.04 (0.87-1.25) 1.06 (0.86-1.32) 
GER 1.22 (0.91-1.63) 1.17 (0.86-1.58) 0.93 (0.62-1.39) 
HAW 0.77 (0.48-1.23) 0.80 (0.48-1.33) 0.68 (0.34-1.33) 
HOP 1.10 (0.89-1.38) 1.16 (0.92-1.45) 1.19 (0.89-1.58) 
MAL 0.96 (0.80-1.15) 0.96 (0.80-1.15) 0.99 (0.80-1.23) 
MAY 0.83 (0.66-1.06) 0.85 (0.66-1.10) 0.91 (0.68-1.24) 
SEA 0.95 (0.83-1.08) 0.94 (0.82-1.08) 0.88 (0.72-1.08) 
STA 0.84 (0.66-1.07) 0.84 (0.65-1.07) 0.92 (0.68-1.24) 
UCI 0.91 (0.73-1.12) 0.86 (0.67-1.11) 0.93 (0.68-1.26) 
UKO 0.86 (0.67-1.09) 0.86 (0.68-1.10) 0.85 (0.62-1.18) 
USC 1.01 (0.85-1.22) 1.08 (0.89-1.31) 1.10 (0.88-1.38) 
Combined 0.99 (0.94-1.05) 0.99 (0.93-1.05) 1.01 (0.94-1.09) 
SiteAll cases,* OR (95% CI)Invasive cases,* OR (95% CI)Serous invasive cases,* OR (95% CI)
AUS 1.14 (0.99-1.31) 1.08 (0.93-1.26) 1.17 (0.98-1.39) 
DOV 1.07 (0.91-1.26) 1.04 (0.87-1.25) 1.06 (0.86-1.32) 
GER 1.22 (0.91-1.63) 1.17 (0.86-1.58) 0.93 (0.62-1.39) 
HAW 0.77 (0.48-1.23) 0.80 (0.48-1.33) 0.68 (0.34-1.33) 
HOP 1.10 (0.89-1.38) 1.16 (0.92-1.45) 1.19 (0.89-1.58) 
MAL 0.96 (0.80-1.15) 0.96 (0.80-1.15) 0.99 (0.80-1.23) 
MAY 0.83 (0.66-1.06) 0.85 (0.66-1.10) 0.91 (0.68-1.24) 
SEA 0.95 (0.83-1.08) 0.94 (0.82-1.08) 0.88 (0.72-1.08) 
STA 0.84 (0.66-1.07) 0.84 (0.65-1.07) 0.92 (0.68-1.24) 
UCI 0.91 (0.73-1.12) 0.86 (0.67-1.11) 0.93 (0.68-1.26) 
UKO 0.86 (0.67-1.09) 0.86 (0.68-1.10) 0.85 (0.62-1.18) 
USC 1.01 (0.85-1.22) 1.08 (0.89-1.31) 1.10 (0.88-1.38) 
Combined 0.99 (0.94-1.05) 0.99 (0.93-1.05) 1.01 (0.94-1.09) 
*

Total cases = 5,877, total invasive cases = 4,774 and total serous invasive cases = 2,583 among 12 OCAC studies, excluding NCO.

ORs are age-adjusted.

For all cases, Cochran's Q = 15.147, P = 0.18, and I2 = 27. For invasive cases, Cochran's Q = 12.173, P = 0.35, and I2 = 10. For serous invasive cases, Cochran's Q = 10.053, P = 0.53, and I2 = 0. Low, moderate, and high levels of heterogeneity correspond to I2 values of 25, 50, and 75, respectively (36).

Figure 1.

Estimated age-adjusted ORs (boxes) and 95% CIs (lines) from unconditional logistic regression models for all cases and controls for each of the 12 OCAC replication studies. The combined estimate is from a fixed-effects model. The size of each box is proportionate to the size of the study.

Figure 1.

Estimated age-adjusted ORs (boxes) and 95% CIs (lines) from unconditional logistic regression models for all cases and controls for each of the 12 OCAC replication studies. The combined estimate is from a fixed-effects model. The size of each box is proportionate to the size of the study.

Close modal

This is the first report to describe a comprehensive assessment of the relationship between common genetic variation in the IL18 gene and ovarian cancer risk among White, non-Hispanic women. One prior small study investigated a single IL18 polymorphism, not in linkage disequilibrium with IL18 rs1834481, and reported null results (19). In the NCO study of 1,536 SNPs from several candidate gene pathways, the IL18 gene showed the strongest evidence for association with epithelial ovarian cancer risk. One of the 11 IL18 SNPs (rs1834481) was significantly associated with epithelial ovarian cancer and uniquely tagged a significant IL18 haplotype, but further studies of this SNP in 12 independent studies by OCAC did not replicate this finding. The most likely explanation for the initial finding is that it represents chance (that is, a type I error). The magnitude of the q value (0.240) for IL18 was moderate, suggesting that the association between IL18 and epithelial ovarian cancer could possibly be a false-positive association. However, it is also important to consider that, although our results suggest that IL18 rs1834481 is not associated with ovarian cancer, we cannot rule out the possibility that other IL18 polymorphisms may be associated with ovarian cancer risk.

When we fitted the fixed- and random-effects models in the replication stage, we did so by first excluding the initial NCO findings. Without NCO, there was no heterogeneity of results and the overall effect of the IL18 SNP was null. With the NCO included, the overall effect of IL18 rs1834481 remained null, but there was significant heterogeneity. Because the NCO study was the first to report a positive finding for IL18 rs1834481, it may have exhibited the “winner's curse phenomenon” (35), whereby the initial finding is often the strongest and most significant, thereby making it statistically different from subsequent studies.

The current study shows the important role of consortia, such as the OCAC, in replication of initial positive findings. Through the OCAC, we were able to refute an initial positive association between a SNP and epithelial ovarian cancer risk via coordinated genotyping and centralized analysis rather than subsequent, potentially conflicting reports from individual studies. Due to the large sample size, in both the number of studies (that is, 13) and the total number of subjects (that is, 15,314), it is unlikely that IL18 rs1834481 is associated with epithelial ovarian cancer. Consortia such as the OCAC fill an important role for achieving the sample size and power needed to detect modest associations with common SNPs and aid in confirmation of initial findings from smaller studies.

No potential conflicts of interest were disclosed.

Grant support: Genotyping was funded by a grant from the Ovarian Cancer Research Fund provided by the family and friends of Kathryn Sladek Smith (Principal Investigator: Andrew Berchuck). U.S. Army Medical Research and Materiel Command grant DAMD17-01-1-0729, Cancer Council Tasmania and Cancer Foundation of Western Australia, and National Health and Medical Research Council of Australia 199600 (AUS); R01CA112523 (DOV); Federal Ministry of Education and Research Programme of Clinical Biomedical Research grant 01 GB 9401 (GER); USPHS grant R01-CA-58598 and NIH Department of Health and Human Services contracts N01-CN-67001 and NO1-CN-25403 (HAW); DAMD 17-02-1-0669 and R01CA095023 (HOP); Mermaid (MAL); NIH grant 1-R01-CA122443 (MAY); NIH grant 1-R01-CA76016 and Department of Defense grant DAMD17-02-1-0666 (NCO); Cancer Research UK (SEA); Roswell Park Alliance and National Cancer Institute grant CA71766 and core grant CA16056 (STA); NIH/National Cancer Institute grants CA-58860 and CA-92044 and Lon V. Smith Foundation grant LVS-39420 (UCI); Cancer Research UK project grant C8804/A7058, Oak Foundation, Eve Appeal, and Department of Health's NIHR Biomedical Research Centers funding scheme (UKO); and California Cancer Research Program grants 00-01389V-20170 and 2110200, USPHS grants CA14089, CA17054, CA61132, CA63464, N01-PC-67010, and R03-CA113148, and California Department of Health Services subcontract 050-E8709 as part of its statewide cancer reporting program (USC). NIH/National Cancer Institute grant T32 CA009330-26 (R.T. Palmieri), Mermaid component of the Eve Appeal (S.J. Ramus), Wellbeing (H. Song), and National Health and Medical Research Council of Australia fellowships (G. Chenevix-Trench and P.M. Webb).

Note: Supplementary data for this article are available at Cancer Epidemiology Biomarkers and Prevention Online (http://cebp.aacrjournals.org/).

D.F. Easton is a principal research fellow of Cancer Research UK. P.D.P. Pharoah is a Cancer Research UK senior clinical research fellow.

The costs of publication of this article were defrayed in part by the payment of page charges. This article must therefore be hereby marked advertisement in accordance with 18 U.S.C. Section 1734 solely to indicate this fact.

Australian Cancer Study/Australian Ovarian Cancer Study: The Australian Ovarian Cancer Study Management Group (D. Bowtell, G. Chenevix-Trench, A. deFazio, D. Gertig, A. Green, and P.M. Webb) gratefully acknowledges the contribution of all the clinical and scientific collaborators.30

The Australian Ovarian Cancer Study and the Australian Cancer Study Management Group (A. Green, P. Parsons, N. Hayward, P.M. Webb, and D. Whiteman) thank all of the project staff, collaborating institutions, and study participants.

We thank all the study participants who contributed to this research.

1
Jemal A, Siegel R, Ward E, Murray T, Xu J, Thun MJ. Cancer statistics, 2007.
CA Cancer J Clin
2007
;
57
:
43
–66.
2
Whittemore AS, Harris R, Itnyre J; Collaborative Ovarian Cancer Group. Characteristics relating to ovarian cancer risk: collaborative analysis of 12 US case-control studies. IV. The pathogenesis of epithelial ovarian cancer.
Am J Epidemiol
1992
;
136
:
1212
–20.
3
Whittemore AS, Harris R, Itnyre J. Characteristics relating to ovarian cancer risk: collaborative analysis of 12 US case-control studies. II. Invasive epithelial ovarian cancers in White women. Collaborative Ovarian Cancer Group.
Am J Epidemiol
1992
;
136
:
1184
–203.
4
Riman T, Nilsson S, Persson IR. Review of epidemiological evidence for reproductive and hormonal factors in relation to the risk of epithelial ovarian malignancies.
Acta Obstet Gynecol Scand
2004
;
83
:
783
–95.
5
Ness RB, Grisso JA, Cottreau C, et al. Factors related to inflammation of the ovarian epithelium and risk of ovarian cancer.
Epidemiology
2000
;
11
:
111
–7.
6
Song H, Ramus SJ, Quaye L, et al. Common variants in mismatch repair genes and risk of invasive ovarian cancer.
Carcinogenesis
2006
;
27
:
2235
–42.
7
Auranen A, Song H, Waterfall C, et al. Polymorphisms in DNA repair genes and epithelial ovarian cancer risk.
Int J Cancer
2005
;
117
:
611
–8.
8
Espey LL. Current status of the hypothesis that mammalian ovulation is comparable to an inflammatory reaction.
Biol Reprod
1994
;
50
:
233
–8.
9
Terranova PF, Rice VM. Review: cytokine involvement in ovarian processes.
Am J Reprod Immunol
1997
;
37
:
50
–63.
10
Behrman HR, Kodaman PH, Preston SL, Gao S. Oxidative stress and the ovary.
J Soc Gynecol Investig
2001
;
8
:
S40
–2.
11
Murdoch WJ, Martinchick JF. Oxidative damage to DNA of ovarian surface epithelial cells affected by ovulation: carcinogenic implication and chemoprevention.
Exp Biol Med Maywood
2004
;
229
:
546
–52.
12
Wang D, Dubois RN. Prostaglandins and cancer.
Gut
2006
;
55
:
115
–22.
13
Grisham MB, Jourd'heuil D, Wink DA. Review article: chronic inflammation and reactive oxygen and nitrogen metabolism—implications in DNA damage and mutagenesis.
Aliment Pharmacol Ther
2000
;
14
Suppl 1:
3
–9.
14
Ristimaki A. Cyclooxygenase 2: from inflammation to carcinogenesis.
Novartis Found Symp
2004
;
256
:
215
–21; discussion 21–6, 59–69.
15
Schildkraut J, Moorman P, Bland A, et al. Cyclin E overexpression in epithelial ovarian cancer characterizes an etiologic subgroup.
Cancer Epidemiol Biomarkers Prev
2008
;
17
:
585
–93.
16
Merritt MA, Green AC, Nagle CM, Webb PM. Talcum powder, chronic pelvic inflammation and NSAIDs in relation to risk of epithelial ovarian cancer.
Int J Cancer
2008
;
122
:
170
–6.
17
Rossing MA, Cushing-Haugen KL, Wicklund KG, Doherty JA, Weiss NS. Menopausal hormone therapy and risk of epithelial ovarian cancer.
Cancer Epidemiol Biomarkers Prev
2007
;
16
:
2548
–56.
18
Royar J, Becher H, Chang-Claude J. Low-dose oral contraceptives: protective effect on ovarian cancer risk.
Int J Cancer
2001
;
95
:
370
–4.
19
Bushley AW, Ferrell R, McDuffie K, et al. Polymorphisms of interleukin (IL)-1α, IL-1β, IL-6, IL-10, and IL-18 and the risk of ovarian cancer.
Gynecol Oncol
2004
;
95
:
672
–9.
20
Pearce CL, Wu AH, Gayther SA, et al. Progesterone receptor variation and risk of ovarian cancer is limited to the invasive endometrioid subtype: results from the Ovarian Cancer Association Consortium pooled analysis.
Br J Cancer
2008
;
98
:
282
–8.
21
Soegaard M, Jensen A, Hogdall E, et al. Different risk factor profiles for mucinous and nonmucinous ovarian cancer: results from the Danish MALOVA study.
Cancer Epidemiol Biomarkers Prev
2007
;
16
:
1160
–6.
22
Sellers TA, Huang Y, Cunningham J, et al. Association of single nucleotide polymorphisms in glycosylation genes with risk of epithelial ovarian cancer.
Cancer Epidemiol Biomarkers Prev
2008
;
17
:
397
–404.
23
McGuire V, Felberg A, Mills M, et al. Relation of contraceptive and reproductive history to ovarian cancer risk in carriers and noncarriers of BRCA1 gene mutations.
Am J Epidemiol
2004
;
160
:
613
–8.
24
Ziogas A, Gildea M, Cohen P, et al. Cancer risk estimates for family members of a population-based family registry for breast and ovarian cancer.
Cancer Epidemiol Biomarkers Prev
2000
;
9
:
103
–11.
25
Ramus S, Vierkant R, Johnatty S, et al. Consortium analysis of 7 candidate SNPs for ovarian cancer.
Int J Cancer
2008
;
123
:
380
–88.
26
Pike MC, Pearce CL, Peters R, Cozen W, Wan P, Wu AH. Hormonal factors and the risk of invasive ovarian cancer: a population-based case-control study.
Fertil Steril
2004
;
82
:
186
–95.
27
Carlson CS, Eberle MA, Rieder MJ, Yi Q, Kruglyak L, Nickerson DA. Selecting a maximally informative set of single-nucleotide polymorphisms for association analyses using linkage disequilibrium.
Am J Hum Genet
2004
;
74
:
106
–20.
28
Storey JD, Tibshirani R. Statistical significance for genomewide studies.
Proc Natl Acad Sci U S A
2003
;
100
:
9440
–5.
29
Raftery AE. Bayesian model selection in social research.
Sociol Methodol
1995
;
25
:
111
–63.
30
R Development Core Team (2008). R: a language and environment for statistical computing. Vienna (Austria): R Foundation for Statistical Computing. ISBN 3-900051-07-0. Available from: http://www.R-project.org.
31
Barrett JC, Fry B, Maller J, Daly MJ. Haploview: analysis and visualization of LD and haplotype maps.
Bioinformatics
2005
;
21
:
263
–5.
32
DerSimonian R, Laird N. Meta-analysis in clinical trials.
Control Clin Trials
1986
;
7
:
177
–88.
33
Cochran W. The combination of estimates from different experiments.
Biometrics
1954
;
10
:
101
–29.
34
Higgins JP, Thompson SG. Quantifying heterogeneity in a meta-analysis.
Stat Med
2002
;
21
:
1539
–58.
35
Kavvoura FK, Ioannidis JP. Methods for meta-analysis in genetic association studies: a review of their potential and pitfalls.
Hum Genet
2008
;
123
:
1
–14.
36
Higgins JP, Thompson SG, Deeks JJ, Altman DG. Measuring inconsistency in meta-analyses.
BMJ
2003
;
327
:
557
–60.

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