Pelvic inflammatory disease and the risk of ovarian cancer: a meta-analysis.

Ovarian cancer is a gynecologic malignancy with a high mortality rate affecting many women worldwide. The World Health Organization (WHO) estimated that the number of ovarian cancer deaths worldwide in 2015 was 162,129 [ 1 ]. In 2017, the expected numbers of new cancer cases and deaths in the United states are 22,440 and 14.080 respectively, and the number of deaths ranks the 5 th of all fatal cancers in women [ 2 ]. Epithelial ovarian cancer makes up 90% of all histologic types [ 3 ]. Borderline ovarian tumors, or tumors with low malignant potential (LMP), are considered as clinical-pathologic features intermediate between benign and malignant with no stromal invasion and account for approximately 10~30% of non-benign ovarian tumors (NBOT) [ 4 ]. Ovarian carcinogenesis is not well understood [ 5 ], and chronic inflammation has been hypothesized as a potential mechanism [ 6 – 8 ]. The ‘inflammation’ hypothesis arises because ovarian cancer has been linked to events and conditions which are related to inflammation and repair (e.g., ovulation, endometriosis) [ 6 ]; while reduced risks have been observed for agents associated with direct anti-inflammatory actions (e.g., aspirin) [ 9 ]. Tubal ligation and hysterectomy are also associated with a reduced risk of ovarian cancer, likely because the direct route of exposure of initially transformed cells to the ovaries are blocked [ 6 ]. The factors released during inflammation and the repeated involvement of the affected ovarian epithelium both play critical roles in the pathological process of ovarian carcinogenesis [ 10 ]. Although historically viewed as arising from ovarian surface epithelium, recent data suggest that many of these tumors, specifically serous tumors, develop from the fibria of the fallopian tube; specifically, that early carcinomas develop in the tube but grow robustly on the ovary [ 11 , 12 ]. Several previous studies indicated that pelvic inflammatory disease (PID) might be a risk factor of ovarian cancer [ 7 , 13 – 15 ]. PID is caused by sexually transmitted infections that ascend from the lower genital tract to produce infections at various sites in the upper genital tract, which include endometritis, salpingitis, pelvic peritonitis, and tubo-ovarian abscess [ 16 ]. Inflammatory stimuli by PID could influence ovarian cancer via a number of possible mechanisms. Chronic inflammation might directly influence the ovarian surface via malignant transformation of epithelial cells, or may facilitate the conversion of premalignant lesions in the fallopian tubes to malignant lesions [ 8 ]. A small meta-analysis of PID and ovarian cancer was conducted in 2011 and included four studies in total [ 14 ]. Among the seven studies published in or before 2011, three of them [ 14 , 17 , 18 ] were not included in the prior meta-analysis [ 14 ]. Since 2011, six additional studies [ 15 , 19 – 23 ] have been published. Furthermore, due to the limited number of studies/participants included in the previous meta-analysis [ 14 ], the potential sources of heterogeneity were not explored. Therefore, we undertook an updated meta-analysis to quantitatively assess the association between PID and ovarian cancer and to explore potential sources of heterogeneity.

We performed a systematic literature search of the PubMed (from 1966 to October 2016), Web of science (up to October 2016) and Embase (from 1980 to September 2016) electronic databases. To ensure that the meta-analysis was based on up-to-date results, we updated the literature search in PubMed on January 12, 2017. The following search terms and/or medical subject headings (MESH) were used: (“pelvic inflammatory” or “pelvic disease” or “adnexitis” or “ovary inflammation” or “oophoritis” or “parametritis” or “salpingitis” or “endometritis” or “inflammatory pelvic” or (“Pelvic inflammatory disease”[mesh])) and (((ovary or ovarian) and (cancer or cancers or carcinoma* or neoplasm* or malignan* ortumour or tumor)) or (“Ovarian Neoplasms”[MESH])). We also reviewed the reference lists of identified original and review articles to search for additional relevant studies. Only those articles published as full-text research papers in English were included. For inclusion, studies had to fulfill the following criteria: (1) the study was a case-control study (including a nested case-control study) or a cohort study; (2) the exposure was PID (including endometritis, salpingitis, pelvic peritonitis, and tubo-ovarian abscess); (3) the outcome was ovarian cancer, including invasive ovarian cancer and/or borderline ovarian tumors; and (4) the study reported odds ratios (OR), relative risks (RR), or hazard ratios (HR), and their corresponding 95% CIs. If multiple reports were published from the same study population, we included the most recent publication. Other pelvic diseases (e.g. endometriosis) were excluded, as they were not the focus of the current meta-analysis. A total of 2,521 articles including potential duplicates across data sources were retrieved via literature search of the PubMed, Web of Science, and Embase databases ( Figure 1 ). After title and abstract scanning, 2,461articles were excluded (meta-analysis and editorial: N = 289; and irrelevant studies, such as animal studies, in vivo studies, or clinical trials: N = 2,172). We reviewed the full texts of the remaining 60 articles. Among them, 48 articles were excluded, as they did not report risk estimates or 95% CIs (N = 43), or were duplicates (N = 5). Thirty-three additional articles were identified with the updated literature search on January 12, 2017, and one new eligible study was included [ 22 ]. A total of 13 studies were included in this meta-analysis, including six cohort studies [ 14 , 15 , 19 , 20 , 22 , 23 ] and seven case-control studies [ 7 , 13 , 17 , 18 , 21 , 24 , 25 ]. Three studies included only invasive ovarian cancer [ 19 , 20 , 24 ], one study included only borderline ovarian tumors [ 23 ], and nine studies included both invasive ovarian cancer and borderline ovarian tumors [ 7 , 13 – 15 , 17 , 18 , 21 , 22 , 25 ]. Of these nine studies, seven reported combined results only [ 13 – 15 , 17 , 18 , 22 , 25 ], one reported results of combined, invasive, and borderline tumors [ 7 ], and one reported results of invasive cancer and borderline tumors separately, but not combined results [ 21 ]. Data were extracted independently by two investigators (ZZ and ZF) and differences were resolved by discussion with a third investigator (SX). Extracted data included the first author’s name, year of publication, country of origin of the studies as a proxy for race, number of participants, mean or median age, duration of follow-up (for cohort studies), ascertainment and category of PID, outcome assessment (diagnosis), ORs or HRs of the outcome and the corresponding 95% CIs for each category of PID, and the covariates adjusted. The quality of each study was assessed by the Newcastle-Ottawa quality assessment scale (NOS) [ 26 ]. This instrument assesses the quality of studies in three aspects ‘selection of cohorts or cases and controls (4 stars)’, ‘comparability of cohorts or cases and controls (2 stars)’, and ‘assessments of outcome (cohort studies) or ascertainment of exposure (case-control studies) (3 stars)’. The quality scores of the included studies ranged from 0 to 9, with 7 to 9 points indicating a high-quality study and 0 to 6 points indicating low-quality. Because ovarian cancer is a rare outcome, ORs were considered acceptable approximations of RR and combined with RRs, resulting in a common estimate of RR [ 27 ]. Summary RR was calculated to quantify the association between PID and ovarian cancer. We transformed the RRs or ORs in each study by using their natural logarithms, and the standard errors (SEs) were calculated from the logarithmic-transformed values and their corresponding 95% CIs. The pooled RR with 95% CI was calculated based on the DerSimonian and Laird method [ 28 ]. Heterogeneity across studies was assessed using I 2 [ 29 ], and high heterogeneity is defined as an I 2 value of greater than 50% [ 30 ]. In the absence of significant heterogeneity, the fixed-effects model was used, and in its presence, the random-effects model was used to estimate the summary RR. Because high heterogeneity was observed in the overall analyses, subgroup analyses were further conducted to explore the potential sources of heterogeneity selected a priori , including race, study design, tumor invasiveness, and adjustment for parity, family history of ovarian cancer, or oral contraceptive use. For the subgroup analysis by tumor invasiveness, five studies were included for invasive ovarian cancer only [ 7 , 19 – 21 , 24 ], three were included for borderline ovarian tumors only [ 7 , 21 , 23 ], and seven studies were included for combined invasive ovarian cancer and borderline ovarian tumors [ 13 – 15 , 17 , 18 , 22 , 25 ]. Although the Risch study [ 7 ] reported results for invasive ovarian cancer, borderline ovarian tumors, and combined, only the results of borderline tumor and invasive cancer were used to calculate the subgroup risk estimates. Publication bias was assessed by visual inspection of funnel plot, and Egger’s linear regression asymmetry test was used to measure the degree of asymmetry of the funnel plot [ 31 , 32 ]. Influence analyses were performed to evaluate the influence of a single study on the overall estimate by excluding one study at a time. Statistical analyses were performed using STATA 12.1 software (College Station, TX, USA).

Table 1 summarizes the characteristics of the meta-analyzed studies. Ten were conducted among primarily Caucasian populations and three among Asian populations. Because age was the most important confounder, all studies were adjusted for age. Additional adjustment for family history of ovarian cancer [ 13 , 19 , 21 , 25 ], use of oral contraceptives [ 7 , 13 , 17 , 18 , 21 , 25 ], education [ 13 , 17 , 18 , 24 ], and menopausal status [ 13 , 18 ] was made in some studies. Except the Lin et al. study, which had a maximum follow-up of 3 years [ 14 ], the mean/median follow-up period was 8.84 years [ 22 ], 17 years [ 19 ], 20.3 years in the inflammation cohort and 18.4 years in the control cohort [ 15 ], and 35 years [ 20 , 23 ] years, respectively, in the six cohort studies. All cohort studies used hospital records, whereas all case-control studies used self-reported PID history collected via interviews or questionnaires ( Table 1 ). The median quality score of the 13 publications was 7, with a range of 6–8 ( Table 2 ). Overall, PID was associated with an increased risk of ovarian cancer (RR1.24, 95% CI 1.06–1.44), with high between-study heterogeneity ( I 2 = 58.8%, p for heterogeneity = 0.003) ( Table 3 and Figure 2 ). No evidence of publication bias was observed in the funnel plot or Egger’s linear regression asymmetry test ( p = 0.743) ( Figure 3 ). Subgroup analyses were performed to explore a priori defined potential source of heterogeneity ( Table 3 ). PID was significantly associated with ovarian cancer among Asians (RR 1.69, 95% CI 1.22, 2.34; I 2 = 0%, p for heterogeneity =0.504) [ 14 , 22 , 24 ]. A lower and marginally significant increased risk was observed in predominantly Caucasian studies (RR 1.18, 95% CI 1.00–1.39; I 2 = 60.7%, p for heterogeneity = 0.005) [ 7 , 13 , 15 , 17 – 21 , 23 , 25 ]. A significant heterogeneity was observed between Asian and Caucasian populations ( p for between-group heterogeneity = 0.054). In the analysis stratified by study design, no a significant difference in risk estimates between cohort and case-control studies was observed ( p for between-group heterogeneity = 0.472). PID was associated with an increased risk of ovarian cancer in both cohort studies (RR 1.32, 95% CI 1.05–1.66; I 2 = 64.7%, p for heterogeneity = 0.015) [ 14 , 15 , 19 , 20 , 22 , 23 ] and case-control studies (RR 1.17; 95% CI 0.93–1.49; I 2 = 57.6%, p for heterogeneity = 0.021) [ 7 , 13 , 17 , 18 , 21 , 24 , 25 ], albeit the association was not statistically significant in case-control studies. PID history was associated with an increased risk of borderline ovarian tumors (RR 1.42, 95% CI 1.25–1.63; I 2 = 0%, p for heterogeneity = 0.756) [ 7 , 21 , 23 ], whereas the association was not significant for invasive ovarian cancer (RR 1.06, 95% CI 0.85–1.32; I 2 = 50%, p for heterogeneity = 0.092) [ 19 – 21 , 24 ]; while the association for combined ovarian cancer and borderline ovarian tumors was elevated, although only marginally statistically significant (RR 1.30, 95% CI 0.99–1.70; I 2 = 38.9%, p for heterogeneity = 0.132) [ 13 – 15 , 17 , 18 , 22 , 25 ] ( p for between-group heterogeneity = 0.083). In addition, a significant positive association between PID and ovarian cancer was observed in studies not adjusted for oral contraceptive use (RR 1.33, 95% CI 1.06–1.67; I 2 = 59.3%, p for heterogeneity = 0.022) [ 14 , 15 , 19 , 20 , 22 – 24 ] ( p for between-group heterogeneity = 0.421). Influence analyses showed that the minimal RR was 1.19 (1.02–1.39) after excluding the Lin study [ 14 ] and the maximal RR was 1.27 (1.07–1.52) after excluding the Rasmussen 2016 invasive ovarian cancer study [ 20 ] ( Table 3 ).

The current meta-analysis quantitatively assessed the association between PID and ovarian cancer. The summary RR supports an association between prior PID diagnosis and increased ovarian cancer risk. Stratified analyses indicated that PID was significantly associated with ovarian cancer risk among Asian women [ 14 , 22 , 24 ], but the risk was lower among Caucasian women [ 7 , 13 , 15 , 17 – 21 , 23 , 25 ]. A significant positive association was observed in cohort but not case-control studies. In addition, PID was significantly associated with an increased risk of borderline ovarian tumors, and marginally associated with invasive ovarian cancer and borderline ovarian tumors combined, but not with invasive ovarian cancer alone. However, this classification was limited and results should be interpreted with caution, given that studies did not systematically report results for each subgroup. In 2011, Lin et al. [ 14 ] conducted a meta-analysis including four case-control studies and found no association between PID and ovarian cancer risk [ 8 ]. In the current meta-analysis, with a total of 13 studies we observed that PID was associated with a 24% increased risk of ovarian cancer. High heterogeneity was found in our meta-analyses. With an additional nine studies, we were able to perform subgroup and influence analyses to further explore the potential sources of heterogeneity. We found that associations varied by race. Further, after excluding the large medical record linkage study by Lin et al. [ 14 ], the risk estimate was attenuated to 1.19 (1.02, 1.39) but still significant in leave-one-out analyses. A recently published large pooled analysis of 13 case-control studies found that PID was associated with an increased risk of borderline ovarian tumors but not ovarian cancer [ 33 ]. Including both cohort and case-control studies in the current meta-analysis, our results are largely consistent with the findings of the pooled analysis, in that we also observed a positive PID-borderline ovarian tumor association, but no association for invasive ovarian cancer only. In addition, we observed an increased risk of combined ovarian cancer and borderline ovarian tumors, which was not examined in the pooled analysis. The increased risk of ovarian cancer associated with PID is biologically and pathologically plausible. In-vitro studies have shown that the cytokines and chemokines released during chronic inflammation can lead to malignant transformation [ 34 ]. In addition, some genes encoding coagulation factors and proteins involved in the inflammatory responses may exhibit tumorigenic functions, thus transforming the affected fallopian tube or ovarian epithelium into malignant cancer [ 35 ]. Pelvic inflammation can, therefore, inactivate cell differentiation and accelerate the development of ovarian tumors. The different risk estimates observed between Caucasian and Asian women may be due to the medical record classification of PID in the study by Lin et al. [ 14 ]. Alternatively, it might be due to the differences in genetics and lifestyle factors, for instance, use of oral contraceptives [ 36 , 37 ] and/or menopausal hormone therapy [ 38 ], between these racial groups. After immigration to the United States, the ovarian cancer risk among Asian women tends to approach that of Caucasian women [ 39 ], suggesting the potentially important roles of lifestyles and environment in ovarian cancer risk. For instance, oral contraceptive pills are protective against ovarian cancer [ 40 ]. Oral contraceptive use in Asian populations might differ notably from that in other, such as Caucasian populations [ 14 ]. In addition, we found that PID was significantly associated with ovarian cancer risk in studies not adjusted for oral contraceptive use. The different mutation rates of genes in these racial groups might also account for the different results [ 41 ], as genes play important roles in the malignant transformation of ovarian epithelium during inflammation. In addition, screening and early treatment of sexually transmitted disease may also reduce severity of PID in the United States and other Western countries. For instance, in the U.S. and European countries, screening for chlamydia is nationwide, and it is considered a reportable disease and is rapidly treated, whereas chlamydia screening is not readily in place at a national level in Asian countries. The increased risk observed among Asian women is based on three studies conducted in Asian populations. In the Shu et al. study [ 24 ], the risk estimate was based on 8 exposed cases and 1 exposed control, resulting in very wide confidence intervals (OR 3.0, 95% CI 0.3–30). The cohort study by Lin et al. [ 14 ] included lower genital tract infections in their definition of PID (ICD-9 code 616). More importantly, they assessed PID exposure during 2004–2005, and in this period women had to have 2 episodes of PID to be counted as exposed, and women were followed until the end of 2006, resulting in a maximum of 3 years of follow-up. This makes a causal association between PID and ovarian cancer less likely. The Shen et al. [ 22 ] study used the same data base as the Lin study, with the particular aim at exploring whether the findings from the Lin study could be reproduced with a longer follow-up [ 22 ]. A lower HR (1.33 95% CI 0.78–2.27) was observed with up to 10 years of follow-up. Although both studies used the same database (Longitudinal Health Insurance Database 2005, LHID2005), because PID cases were diagnosed in different periods (January 1 st , 2004 and December 31 st , 2005 in the Lin study, and January 1 st , 2000 and December 31 st , 2002 in the Shen study), we included both studies in the meta-analysis and observed a significant positive association among Asian women. However, given the limitations of these studies, the association between PID and ovarian cancer among Asian women should be interpreted with caution, and evaluated in future studies. In the analysis stratified by study design, a significantly elevated risk was observed in cohort studies [ 14 , 15 , 19 , 20 , 22 , 23 ], but not in case-control studies [ 7 , 13 , 17 , 18 , 21 , 24 , 25 ]. It should be noted that all cohort studies used hospital records, whereas all case-control studies used interviews or questionnaires for PID assessment. In the cohort studies using hospital records, the PID status was ascertained and confirmed by diagnosis and generally speaking was more objective and accurate than the self-reported assessment methods. Misclassification of PID diagnosis is not entirely avoidable in observational studies, as the diagnostic criteria may have changed over time. In addition, a possibly more important reason for misclassification of PID status is that PID is an extremely difficult exposure to assess, with a suspected large proportion of cases being subclinical and therefore not captured by neither medical records nor self-reported history. However, differential recall between cases and non-cases may be less likely in investigations assessing the PID ovarian cancer association even in case-control studies, as the public awareness of PID and its potential association with ovarian cancer risk may not be high. Thus, the non-differential recall may have attenuated the association further towards the null in case-control studies. Additional large studies should be conducted in the future to further elucidate this association. In the circumstances that a case-control study is the only feasible study design, hospital records should be used for objective and accurate PID assessment method. A significant positive association was observed in the studies which were not adjusted for parity and in those studies which were not adjusted for family history of ovarian cancer. Confounding by parity and/or family history of ovarian cancer needs to be considered carefully in analyses of PID and ovarian cancer risk. Parity does not likely cause PID, but PID can influence subfertility and it is well documented that severe PID can affect tubal patency. Thus, parity is on the causal pathway between PID and ovarian cancer and can only serve as a potential mediator of the PID-ovarian cancer association. It is unlikely that family history of ovarian cancer predisposes women to PID; therefore it also does not satisfy the traditional definition of a potential confounding factor, so adjustment should be interpreted with caution. The meta-analysis results presented in the current manuscript suggest that inappropriate adjustment for these factors may attenuate or even change the direction of the association. The more appropriate estimates of the true association between PID and ovarian cancer are likely the studies not adjusting for these factors. The current meta-analysis has several potential limitations. First, PID is a broad category which includes different types of inflammatory diseases (endometritis, salpingitis, pelvic peritonitis, and tuba-ovarian abscess). These diseases may play different roles in ovarian carcinogenesis and correspond to different levels of ovarian tumor risk. Unfortunately, with the exception of one study which specifically focused on salpingitis as the PID exposure [ 13 ], the other studies grouped the different types of diseases in the broad category of PID or pelvic inflammation, limiting our ability to assess the associations between different types of PID and ovarian cancer risk. Further, the cause of PID (e.g. type of sexually transmitted infection) and details of PID exposure, for instance, number of PID episodes, age at first PID, or time since first PID, were not available in most studies. In addition, we were not able to perform an analysis stratified by histologic subtype of ovarian cancer, as only three studies reported histotype-specific associations (one on combined invasive cancer and borderline ovarian tumors [ 17 ], one on invasive ovarian cancer [ 20 ], and one on borderline ovarian tumors [ 23 ]), although ovarian cancer is a heterogeneous disease with different risk factor profiles[ 20 ]. Second, of the total 13 studies included, the six (less than half of the total studies included) cohort studies [ 14 , 15 , 19 , 20 , 22 , 23 ] assessed PID status based on documented medical records, whereas all the seven case-control studies [ 7 , 13 , 17 , 18 , 21 , 24 , 25 ] used self-reported PID history. We did observe a statistically significant and higher risk in cohort studies than in case-control studies. Considering the concern of recall bias in case-control studies, additional large cohort studies with long follow-up and objective and accurate PID assessment methods and adequate adjustment of important confounders should be conducted to provide valid estimates to better quantify the magnitude of the PID-ovarian cancer association. In conclusion, this meta-analysis supports a positive association between PID and increased risk of ovarian cancer. The association was more pronounced among Asian than Caucasian women. Sexually transmitted infections need to be treated early to avoid potential adverse sequelae of PID. The association also seemed to be most apparent in analyses of borderline ovarian tumors. However, given that the studies reporting combined results cannot be disentangled to determine the proportion related to borderline vs. invasive tumors, the results by tumor invasiveness should be interpreted with caution. Future large and well-designed cohort studies with objective and accurate PID assessment and adequate adjustment for important confounders need to be conducted to confirm the stronger association observed among Asian women and to determine if differential associations exist for borderline vs. invasive tumors.