We therefore carried out a Relevant studies were searched from PubMed database, CNKI, WANFANG DATA and VIP INFORMATION up to January 2012, with no language restrictions. References listed from selected papers were also reviewed. We included studies that reported the estimates of relative risks (RRs) with 95% confidence intervals (CIs) for the association between menstrual and reproductive factors and lung cancer risk. The pooled RRs were calculated after the heterogeneity test with the software Stata 11, and publication bias and sensitivity were evaluated at the same time.meta-analysis. Older age at menarche in North America women  was associated with a significant decreased risk of lung cancer. Longer length of menstrual cycle was also associated with decreased lung cancer risk . Other exposures were not significantly associated.Twenty-five articles, representing 24 independent studies, were included in this Our analysis provides evidence of the hypothesis that female sex hormones influence the risk of lung cancer in women, yet additional studies are warranted to extend this finding and to clarify the underlying mechanisms. The incidence of lung cancer is increasing in women, in contrast to that seen in men. Interestingly, the proportion of lung cancer cases in females attributable to smoking is approximately half of that seen in males. In addition, there is new evidence suggesting that estrogen receptors (ERs) ERα and ERβ have been detected on lung cancer cells in females, 5. Laboratory evidence proves that estrogen acts as a promoter for lung adenocarcinoma in a mouse model based on genetic alterations that are relevant to the human condition. These ﬁndings have suggested a hypothesis that female sex hormones, and consequently menstrual and reproductive factors, may play an important role in lung carcinogenesis.Lung cancer remains the leading cause of death from cancer among women in the United Statesmeta-analysis of these studies.Results from epidemiological studies, however, have been controversial. To evaluate the relationship between risk of lung cancer and menstrual and reproductive factors, we conducted a meta-analysis in accordance with the meta-analysis of Observational Studies in Epidemiology guidelines. We conducted a systematic literature search of the PubMed database, CNKI, WANFANG DATA and VIP INFORMATION through January 2012 by using the following search strategy:  AND (lung cancer OR lung neoplasms OR pulmonary cancer OR pulmonary neoplasms), with no restrictions. Reference lists of the chosen papers were also reviewed for other potential articles that may have been missed in the database search. We did not contact the authors of the primary studies to request addition information.We attempted to report this Studies were included in our analysis if they met the following criteria: 1) the study design was a case-control study or a cohort study; 2) the endpoint of cohort study was incidence of lung cancer; 3) the risk estimate of lung cancer related to menstrual or reproductive factors and the corresponding 95% confidence intervals (CIs) were reported; studies without existing RRs were also eligible if the relevant data presented in the study we can use to calculate crude RRs instead; 4) of the studies with the same or overlapping population published in more than one study, we included the most recent ones.Two authors independently reviewed the articles and extracted the data, any disagreement was resolved by discussion. Information was recorded for each study as follows: last name of the first author, year of publication, study location, type of study design, length of follow-up (if applicable), total person-years of observation (if applicable), studied outcome, exposure variables and categories, number of lung cancer cases, number of controls, cohort size, age range (if applicable), menopausal status of the participants (if applicable), smoking status of the participants (if applicable), adjusted RR estimates from multivariable model with corresponding 95%CIs and statistical adjustment for potential confounders of interest., including characteristics of study populations, assessments of exposure and outcome and confounding variables controlled.Instead of providing aggregate scores, we assessed the quality of individual studies by recording the key components of study designsmeta-analysis of the comparison of the highest versus the lowest category in each study. In addition, exposures to OC use and HRT use were analyzed as dichotomous variables. Because some studies reported associations for different durations of OC use and HRT use in comparison with never use, we pooled those RRs to evaluate the total effect for ever versus never use.In this reanalysis, the odds ratio (OR) and the hazard ratio (HR) were directly considered as RR. The categories of menstrual and reproductive exposure measures varied across studies, so we performed this Pooled risk estimates were calculated for exposure variables that were reported in at least five studies, which included age at menarche, length of menstrual cycle, number of pregnancies, parity (number of live births), age at first live birth, age at menopause, oral contraceptive (OC) use and hormone replacement therapy (HRT) use.Other menstrual and reproductive variables reported in less than five studies included menopause status, other hormones, type of surgery, number of difficult labour, positive history of hysterectomy; miscarriage; spont; abortion, breast feeding, period length, menstrual regularity, quantity of menstrual flow, mental tension or pains related to menses, age at first use OC, age at first use HRT, years of menstruation, time since first use OC, time since last use OC, type of HRT, lifetime estrogen dose from HRT, intrauterine device use, age at last birth, estrogen plus progestin pills, calendar year of first use OC, HRT and number of menstrual cycles up to the reference date.Q statistic at the P < 0.05 level of significance. We also calculated the I2 statistic, a quantitative measure of inconsistency across studies, which was classified as low (25%), moderate (50%) and high (50%). RRs from different studies were pooled using the fixed effect model and the random effect model based on the Mantel-Haenszel method and the Dersimonian and Laird method respectively. If there is no heterogeneity between studies, we used a fixed effect model; otherwise, we adopt the random effect model.Heterogeneity test was performed with the use of meta-analysis RR by sequentially omitting each one before pooling study-specific RRs.Subgroup analyses according to geographic region and study design were carried out to assess the potential impacts on the association. In addition, we conducted a sensitivity analysis to examine the influence of individual studies on the overall Begg's funnel plots and formally by Egger's linear regression test, 11. All statistical analyses were done using STATA version 11.0 software. A P value < 0.05 was considered statistically significant and all statistical tests were two-sided.Potential publication bias was assessed by visual inspection of et al, Zheng et al, Dorjgochoo et al and Liu et al because they provided insufficient data. Articles by Zhong et al, Qin et al, Yin et al, and Fang et al were excluded because only point estimates and 95%CI were reported without categorical variables. We further excluded one study which studied on the survival of lung cancer. Two articles reported risk estimates based on the partially overlapping Chinese population, 47, so we extracted data from Zheng et al, the more recent reference, for all menstrual and reproductive variables. Two articles reported partially overlapping data from the Czech Women's Lung Cancer Study, 48, however, Zatloukal et al reported the results separately by cell types of lung cancer, so we extracted all menstrual and reproductive data from Kubik et al. Two articles based on the China Gansu Province population database overlapped, 49, so data from the more recent reference were used. One report analyzed unpublished data collected in a long-standing hospital-based case-control study, which materials and methods were described in detail by Wynder et al, so the information of study was extracted from Wynder et al. Seow et al reported the results separately by smoking status, so we extracted two data sets from the study. Two articles reported partially overlapping data from the prospective Shanghai Women's Health Study, 39, we extracted data from Chen et al, the more recent reference, for all menstrual and reproductive variables except number of children and age at first live birth, which were only available from Weiss et al. Two articles based on the population from Fujian reported overlapping results on age at menarche, 43, so those results were extracted from the more recent reference, other related variables were extracted from Chen et al. Therefore, a total of 25 articles (18 written in English and 7 in Chinese), representing 24 independent studies, were included in this meta-analysis. A flow chart showing the study selection process is presented in Fig 1.The literature search initially identified 926 publications from PubMed, CNKI, WANFANG and VIP INFORMATION; most were excluded because they were reviews or because the exposure or endpoint was not relevant to our analysis. After scanning, 33 were retrieved for further evaluation that also let to identification of 5 more articles from their collective references. Thus, 38 articles reported associations of at least one menstrual or reproductive variable with lung cancer risk. However, we excluded the articles by Fan Tab 1. The 25 studies were published between 1988 and 2012. Of them, 13 studies had been carried out in Asia, 9 in North America and 3 in Europe. Eighteen studies reported case-control comparisons and 7 were analyses of cohorts.The characteristics of the included studies are presented in -44, 46. However, Zheng et al, Wu-Williams et al, Zhou et al, Seow et al and Matsuo et al used the oldest age at menarche as the referent category, hence these 6 adjusted RRs could not be pooled with others comparing the highest versus the lowest categories, so we calculated crude RRs according to the number of cases and controls, then we used them instead. Furthermore, we excluded one study because it did not provide the required data.Nineteen studies examined the relationship between lung cancer risk and age at menarcheI2=65.2%), but in subgroup analysis according to geographic region we found a significant decreased risk of lung cancer associated with older age at menarche in North America women .Study-specific RRs for the oldest age at menarche as compared with the youngest age ranged from 0.345 to 3.158, and the pooled RR was 0.93 (95%CI: 0.79-1.10). The RR Fig 3 represents a forest plot of the effect size distribution for the seven literatures-29, 31 that studied on length of menstrual cycle. Zheng et al and Zhou et al used the longest length of menstrual cycle as the referent category, so we calculated and used crude RRs instead. Study-specific RRs for the longest versus the shortest length of menstrual cycle ranged from 0.46 to 0.91. The combined RR suggested a significant inverse association with a 28% decreased risk of lung cancer and low between-study heterogeneity., 39, 40, among which two crude RRs, 36 were calculated according to the number of cases and controls. Study-specific RRs for highest number of pregnancies as compared with the lowest ranged from 0.69 to 1.683. The pooled RR was 1.10 (95%CI: 0.91-1.34), with low heterogeneity among the studies.Associations of lung caner risk with number of pregnancies were suggested in seven studies, 44-46 provided information on parity, of which a crude RR was calculated based on the number of cases and controls, with study-specific RRs for highest number of live births in comparison with the lowest ranging from 0.39 to 4.744, and the summary RR was 0.91 . In the eighteen studies, half studies, 45, 46 used the nulliparous women as the reference group, the pooled RR was 0.93 , whereas only three studies, 42, 44 used a parous comparison group (1-2 children) so we did not estimate the risk, others were not clearly classified.Eighteen studies, 44-46, including two crude RRs, 42 instead, and ranged from 0.49 to 1.50. The combined RR was 1.03 (95%CI: 0.88-1.21), with moderate heterogeneity across studies.Risk estimates for oldest versus youngest age at first live birth were reported in 11 studies, 44, 46 reported the relationship between age at menopause and lung cancer risk, including five crude RRs, 44, 46 calculated according to the number of cases and controls. However, two studies were excluded because one used not yet present as the referent category and the other did not provided required data. Study-specific RRs for the oldest age at menopause as compared with the youngest age ranged from 0.39 to 1.7. The pooled RR was 0.91 (95%CI: 0.70-1.17), with high heterogeneity across studies.Sixteen studies, 42, 44 reported the risk estimates for ever versus never OC use. What's more, the studies by Wu et al and Seow et al provided 2 RRs depending on duration of use, which we pooled to acquire overall RRs for ever use of 0.604 (95%CI: 0.273-1.336) for Wu et al and 0.854 (95%CI: 0.615-1.184) for Seow et al. The pooled RR of lung cancer for ever users of OC as compared with never users was 0.97 (95%CI: 0.89-1.06), with modern heterogeneity among the studies.Eight studies, 41, 44, but one study was excluded because OC use was included in HRT use. In addition, the studies by Wu et al, Baik et al and Brinton et al represented 2 RRs depending on duration of use, which we calculated to obtain overall RRs for ever use of 1.074 (95%CI: 0.749-1.539) for Wu et al, 0.948 (95%CI: 0.859-1.046) for Baik et al and 0.934 (95%CI: 0.874-0.998) for Brinton et al. The pooled RR of lung cancer for ever users versus never users of HRT was 0.99 (95%CI: 0.91-1.08), with modern heterogeneity among the studies.Risk estimates for ever versus never HRT use were reported in eight studiesmeta-analysis was omitted at a time to reflect the influence of the individual data set to the pooled RRs, and the corresponding combined RRs were not materially altered , suggesting that our results were stable and reliable.A single study involved in the Begg's funnel plot was conducted to assess the publication bias of studies . Then, the Egger's test was performed to provide statistical evidence of funnel plot symmetry. The P values for Egger's test were greater than 0.05 for all exposure variables with the exception of age at menopause  and HRT use , the results indicated publication bias for age at menopause and HRT use.The meta-analysis identified that decreased lung cancer risks were prone to present in women with longer length of menstrual cycle. We also found that age at menarche of North America women was inversely associated with lung cancer risk. Other six factors did not appear to be strongly associated with risk of this tumor. In summary, these findings support the hypothesis that estrogen exposure has an effect on the risk of lung cancer in women.Several menstrual and reproductive factors have been suggested related to lung cancer risk, however, many of these results are inconsistent. There is a common point that women are more likely to be diagnosed with lung adenocarcinoma and non-small cell lung cancer (NSCLC). Our , 51; 2) hydroxylated estrogen metabolites can undergo redox cycling to generate free radicals, which cause DNA damage and lead to carcinogenic mutations; 3) estrogens can directly stimulate the transcription of estrogen-responsive genes in the nucleus of lung cells and can also transactivate growth-factor-signaling pathways, such as the epidermal growth factor receptor (EGFR) pathway, which was involved in NSCLC growth, protection from apoptosis, and angiogenesis-55; moreover, 4) estradiol (E2) can enhance the expression of midkine (MK) protein and E2 increased MK mRNA expression in lung adenocarcinoma cells. Both estrogen and MK can induce NSCLC epithelial-mesenchymal transition, which plays an important step in the migration of lung tumor cells.Shorter length of menstrual cycle indicated an overall increase in the period of unopposed estrogen exposure, and younger age at menarche implied more menstrual cycles over the lifetime and hence longer periods of estrogen exposure in total. Women who undergo shorter length of menstrual cycle and younger age at menarche may have an increased risk of lung cancer, possibly due to more cumulative exposure to endogenous estrogen, which may be involved in the etiology of this disease. There are several lines of evidence that estrogens may promote lung tumorigenesis: 1) estrogens can exert their biological effect through two ER subtypes, ERα and ERβ, particularly ERβ, which promotes estrogen-dependent growth of lung cancer cellset al found that short length of menstrual cycle (< 27 days) were significantly more common with increasing number of variant A2 alleles. The A2 allele was thought to enhance the transcriptional activity of the CYP17 gene leading to elevated levels of estrogen-60, which may increase the risk of lung cancer.Furthermore, Henningson OC use and HRT use, as surrogates of exogenous sex hormonal exposure, seemed not to have a strong impact on lung cancer risk. This may suggest that endogenous and exogenous sex hormone play different roles in lung tumorigenesis, yet further researches using larger study populations are needed to confirm this assumption.meta-analysis. Some evidence of heterogeneity was observed throughout our study. This was partially owing to the following facts: the studies we included focused on different types of design, most of them were case-control studies; studies we used were conducted in different geographic regions, mostly Asia and North America, where people share little in the field of genetic background, lifestyle, and lung cancer incidence; and the ranges of exposure variables in most studies were inconsistent. On this occasion, subgroup analysis was carried out to explain the heterogeneity. As a result, we found that differences in geographic region might contribute to the heterogeneity between studies.Heterogeneity is often a concern in a Egger's test suggested little evidence of publication bias in our meta-analysis. We cannot preclude the possibility, as with any meta-analysis, that other unpublished studies may have been missed during our literature search. Meanwhile, we could hardly found the articles written in authors' mother tongue. Moreover, studies with null effects were less published than those with positive ones, which made it different for us to obtain.meta-analysis should be considered. First, our analysis was limited by the inconsistent categorization of the exposure variables, especially those with more than two strata. However, all adjusted RRs were estimated on the basis of the highest versus the lowest category of the exposure variables, and the wide ranges of exposure variables probably reduced this bias. Second, residual confounders were always concerned in observational studies. Although we used the reported multivariable adjusted RRs where available, we still could not exclude the probability that other unmeasured factors have influenced the real relationship. Nonetheless, our study had a noteworthy strength. As individual study had insufficient statistical power, our meta-analysis of 24 studies involving a large number of participants enhanced the power to detect significant associations and provided more reliable estimates. Moreover, our results are consistent with the hypothesis that estrogen exposure may increase the risk of lung cancer in women, but the mechanisms involved are likely to be complex. It is clear that further studies, both mechanistic and epidemiologic, are warranted in this area. Our findings provide further evidence on the public health with respect to the lung cancer prevention in women.Potential limitations of our On balance, older age at menarche in North America women  was associated with a significant decreased risk of lung cancer. Longer length of menstrual cycle was also associated with decreased lung cancer risk . The other six exposures were not significantly associated. More investigations in large and well-designed studies are needed to extend these findings and to clarify the underlying mechanisms.This study was supported by grants No.81102194 from National Natural Science Foundation of China, No.LS2010168 from Liaoning Provincial Department of Education, and grant No.00726 from China Medical Board. The authors are most grateful to all the participants in this study.No competing financial interests exist for any of the authors.


AR) gene encode a polyglutamine tract in AR N-terminal transactivation domain. Studies have been conducted to evaluate the effect of CAG repeat length on male infertility, which have yielded contradictory results. This study aimed to explore the number of AR-CAG repeats in 150 fertile controls and 150 idiopathic infertile men, divided into four azoospermia, oligozoospermia, asthenozoospermia, and teratozoospermia subgroups. In addition, a meta-analysis was conducted based on previous studies to assess the association of the mentioned variation with male infertility in recent years. Polymerase chain reaction (PCR) targeting followed by an electrophoresis on polyacrylamide gel was used for AR-CAG genotype detecting. Moreover, a systematic search was performed in PubMed, Web of Science, Science Direct, and Google Scholar databases to collect eligible studies for meta-analysis purpose. According to the results, a significant association was observed between increased length of AR-CAG polymorphism and male infertility (p< 0.0001). Furthermore, there were similar significant associations in the azoospermia (p= 0.048), asthenozoospermia (p= 0.013) and teratozoospermia (p= 0.002) subgroups. In addition, meta-analysis on forty studies showed a significant association between AR-CAG polymorphism in the overall analysis  and the Caucasian subgroup . Our results elucidated that long stretches of CAG repeat might lead to AR dysfunction, contributing to male infertility especially in the Caucasian population.CAG trinucleotide repeats in androgen receptor ( AR gene contains eight exons and seven introns, locating on chromosome Xq11-12 . AR andal., 2016). The usal., 2017; Casellaal., 2012; Andersek+ tubes from all participants and preserved at -20 ºC. An informed written consent was obtained from all the participants prior to the research and this study was approved by the Medical Research Ethics Committee of the Kashan University of Medical Sciences (Reference no. IR.KAUMS.REC.1396.24).In this case-control study, the blood samples of 150 idiopathic infertile men and 150 age-matched healthy fertile control men were prepared. All infertile samples were enrolled from the reproduction clinic of Beheshti hospital affiliated with Kashan University of Medical Sciences . An accurate medical and reproductive history was taken from the infertile subjects through an interview, and all participants were examined for diseases such as orchitis, varicocele, maldescensus testis, immune, obstruction of vas deferens, infectious abnormalities, diabetes mellitus, drug abuse, abnormal hormones profile , abnormal karyotype, and Y chromosome microdeletions. Furthermore, the participants were screened for genetic and familial diseases. Therefore, subjects with any of the above-mentioned disorders were excluded from the study. The subjects in the control group were also randomly selected from fertile men without any history of infertility, who had a minimum one child and completely normal sperm parameters. Finally, 2 ml peripheral blood samples were collected in impregnated EDTATM genomic blood DNA extraction kit  and stored at -20 °C for future applications. Our researchers used the PCR technique for genotyping of the AR-CAG polymorphism. Sense (5'-CCAGAATCTGTTCCAGAGCGTG-3') and antisense (5'- GCTGTGAAGGTTGCTGTTCCTC-3') primers were designed surrounding the polymorphic region by Oligo7 software. In this study, the PCR was performed in a 20 µl total volume containing 10 µl 2X Taq PreMix, 0.35 µM of each primer, and 40 ng DNA template. A thermocycling protocol was carried out in peqSTAR thermal cycler  with the following program: the initial denaturation at 94 °C for five minutes, which continued by 40 repetitive cycles including the denaturation step at 94 °C for 45 seconds, annealing step at 57.7 °C for 45 seconds, extension step at 72 °C for 45 seconds, and final extension at 72 °C for five minutes. All PCR reagents were purchased from CinnaGen Company . For rapid and accurate assessment of CAG repeat numbers, the DNA fragments were run on 8 % polyacrylamide gel and were visualized by AgNO3 staining protocol. After taking photos of the gels, Gel Analyzer 2010a software was used for AR-CAG repeat analyzing  up to August 2018, using the keywords of "male infertility", "androgen receptor", "CAG repeat", and "polymorphism". The selection criteria for publications were as follows: all reports must be on the association between CAG repeat length and male infertility in a case-control design; the case group should be included idiopathic infertile males with insufficient sperm parameters according to the criteria by the World Health Organization (WHO) in 1999 or earliFirst author's name; publication year; country of origin; ethnicity; the number of samples, the mean and standard deviation (SD) of CAG repeat length for cases and controls, and association result. The ethnicity of publications was categorized into the two Asian and Caucasian groups  and 95 % confidence interval (CI). Moreover, Cochran's Q statistic and I2 index were used to evaluate the heterogeneity. In this regard, Pheterogenity less than 0.1 was considered statistically significant. Furthermore, a random-effect model was employed when a true heterogeneity was observed; otherwise, a random-effect model was used. In addition, possible publication bias was analyzed by Begg's funnel plot and Egger test. Statistical analyses for case-control trial and meta-analysis were performed in SPSS version 20 and MedCalc statistical software, respectively.In this case-control study, the mean value of various CAG repeats of infertile patients was compared to those infertile controls using the two-sample independent t-test. For subgroup analysis, one-way ANOVA test was used as appropriate. In addition, a p-value (p) were calculated, results of which are summarized in Table 2p< 0.0001; Figure 2p= 0.048), asthenozoospermia (p= 0.013), and teratozoospermia (p= 0.002) subgroups.To estimate androgen receptor CAG repeats length distribution in cases and controls, the mean (M), standard deviation (SD), and Figure 4p< 0.001; Figure 5Ap= 0.008; Figure 5Bp= 0.228) and African  populations. A true heterogeneity was observed in overall analysis , which remained after stratification analysis in the Caucasian , Asian , and African  subgroups. Moreover, there was no significant publication bias in overall (PEgger= 0.279) and stratified  meta-analysis . It is AR-CAG variation with the risk of male infertility. One of the hazardous regions on exon 1 of the human AR gene coding stretches of identical glutamine residue with polymorphic length variation . Intratal., 2017). Changeal., 2017). Howeveal., 2017), there al., 2017; Buchanaal., 2017). In addal., 2017). It is al., 2017; Nejati al., 2017; Zamani-al., 2017; Noureddal., 2017). One of the major drawbacks of the present study was lack of evaluation of the gene-gene and gene-environmental interactions. Another limitation was restricting the search procedure in English in the meta-analysis, which might have led to language bias. Furthermore, there was no access to original data  from the included studies to adjust our data to them.Mohammad Karimian and Hossein Nikzad (hnikzad10@gmail) contributed equally as corresponding authors.The authors declare that there is no conflict of interest regarding the publication of this paper.This study was supported by Grants from the Kashan University of Medical Sciences (No. 96023).


Recently, laparoscopic appendectomies (LAs) have been widely performed instead of open appendectomies (OAs) during pregnancy. However, concerns about the safety of LA during pregnancy remain. This systematic review and meta-analysis aimed to evaluate the current evidence relating to the safety of LA versus OA for suspected appendicitis during pregnancy.Comprehensive literature searches were conducted using the PubMed, EMBASE, and Cochrane Library databases to identify articles describing LA versus OA in pregnancy, without restrictions regarding the publication date. The primary endpoints were fetal loss and preterm delivery.P = 0.581). A significant difference was not evident between LA and OA with respect to preterm delivery , a result that did not change following the sensitivity analysis. The patients who underwent LA had shorter hospital stays  and a lower wound infection risk  compared with those who underwent OA.After screening 801 studies, 22 comparative cohort studies were included in the analysis, which involved 4694 women, of whom 905 underwent LAs and 3789 underwent OAs. Fetal loss was significantly higher among those who underwent LAs compared with those who underwent OAs, and the pooled odds ratio (OR) was 1.72  without heterogeneity. The sensitivity analysis showed that the effect size was influenced by one of the studies, because its removal resulted in there being no significant difference between LA and OA with respect to the risk of fetal loss (OR 1.163, 95% CI: 0.68–1.99; It is not reasonable to conclude that LA in pregnant women might be associated with a greater risk of fetal loss. The difference between LA and OA with respect to preterm delivery was not significant. Acute appendicitis is the most common nonobstetric surgical problem that occurs during pregnancy, and its incidence varies widely, with rates ranging from 1.8 to 41 per 10,000 pregnancies . The incAppendicitis during pregnancy has been reported to be associated with poor pregnancy outcomes, including fetal loss, preterm delivery, and perinatal morbidity and mortality . Fetal lOpen appendectomy (OA) has been performed on patients with acute appendicitis of both sexes and of all ages, including pregnant women. Moreover, laparoscopic appendectomy (LA) has also become a standard procedure for acute appendicitis since it was first performed in 1983 . AlthougThis systematic review was conducted and is reported according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines .Three major electronic medical databases, namely, PubMed, the Cochrane Central Register of Controlled Trials, and EMBASE, were comprehensively searched to find suitable studies using the following search terms: “pregnancy”, “pregnant women”, “appendicitis”, “appendectomy”, and “laparoscopy”. We restricted the searches to studies that were conducted on human subjects and those that were published in English, but there was no restriction regarding the publication date.The present review included any comparative studies that compared the outcomes from LA and OA for appendicitis in pregnant patients. Those studies with at least one pregnancy outcome, for example, fetal loss, preterm delivery, birth weight, or the Apgar score, or one surgical outcome, for example, the wound infection rate, the intra-abdominal abscess rate, the operative time, or the length of stay (LOS), were included. Case series, review articles, and articles written in languages other than English were excluded from this review. When there were duplicate publications, the study with the largest number of subjects was selected.The titles and the abstracts of the extracted studies were reviewed independently by two researchers (SHL and JYL). The complete manuscripts were reviewed if the abstracts did not provide enough information to indicate suitability for the study. The studies were finally included in the systematic review and meta-analysis after full-text evaluations were performed independently. Any discrepancies were resolved through discussion.This study’s primary endpoints were the pregnancy outcomes, including fetal loss and preterm delivery. The data extracted that included the publications’ general data , the studies’ characteristics , the baseline characteristics of the studies’ populations , the pregnancy outcomes , and the surgical outcomes , were summarized and analyzed. All of the data were cross-checked independently by two authors (SHL and JYL).P < 0.10, and an I2 statistic with a value of ≥50% was considered to indicate substantial heterogeneity. A fixed effects model was used to pool the results when heterogeneity was not suspected, otherwise, a random-effects model was used. Publication bias was evaluated by assessing funnel plot symmetry. A sensitivity analysis, which involved repeating the sequential pooling outcomes while excluding each study in turn, was performed to evaluate whether the overall results were robust in relation to the excluded studies. Meta-regression analyses were performed to examine whether the primary outcomes were associated with other characteristics of the studies, for example, the publication year, complicated appendicitis, the gestational age, the pregnancy trimester, or a negative appendectomy.The statistical analyses were performed using Review Manager, version 5.3  and Comprehensive Meta-Analysis, version 2 . For the continuous outcomes, the results were pooled using the inverse-variance method and the mean differences (MDs) and their 95% confidence intervals (CIs) were calculated. For the dichotomous outcomes, the pooled odds ratios (ORs) and their 95% CIs were calculated using the Mantel-Haenszel method. The heterogeneity of the effect size across the studies was tested using Cochran’s Q test, with the significance level set at The study selection flow diagram is shown in Fig. n = 9) [n = 4) [n = 3) [n = 2) [n = 1) [The characteristics of the included studies are presented in Table n = 9) , 37–39,  [n = 4) , 22, 23, [n = 3) , 36, Tur [n = 2) , 31, Chi [n = 1) , India ( [n = 1) , Netherl [n = 1) , and Tun [n = 1) . The pat [n = 1) . The neg [n = 1) , 31–39,  [n = 1) , 35–38.  [n = 1) , 35, and [n = 1) , 25, 35.P = 0.89; I2 = 0%)   in the fixed effects model (81) Fig. .Fig. 2FoP = 0.92; I2 = 0%)  in the fixed effects model (0%) Fig. . Moreoven = 409) did not show a significant MD between LA and OA  in the fixed effects model . The LA and the OA groups were compared in relation to the Apgar scores in five studies . The data were heterogeneous , and the unstandardized MDs were 0.12  for the Apgar score at 1 min and − 0.02  for the Apgar score at 5 min, indicating that there were no significant differences between the groups with respect to the Apgar scores.The meta-analyses of the other pregnancy outcomes, for example, birth weight and the Apgar score, are shown in Table P = 0.005). Seventeen of the studies reported the LOS [P = 0.001), but the analysis showed heterogeneity .Data describing the wound infection and intra-abdominal abscess rates, the operative times, and the LOS were pooled across the studies Table . Nine of the LOS , 36, 38, the LOS , 31, 35. the LOS , 36, 38, the LOS , 24; the the LOS  defined P = 0.42) or the intra-abdominal abscess rate  [Twelve of the studies reported the operative times , 36, 38, = 0.58) , 25, 30.Meta-regression analyses were conducted to evaluate whether the year in which the study was published, the presence of complicated appendicitis, the gestational age at surgery, the trimester, or the presence of a negative appendectomy influenced the meta-analysis outcomes, for example, fetal loss and preterm delivery. None of the meta-regression analyses were statistically significant Table . The metThe funnel plots for fetal loss and preterm delivery are presented in Fig. The findings from this systematic review and meta-analysis showed that it is not reasonable to conclude that LA during pregnancy might be associated with a greater risk of fetal loss. Indeed, no significant differences were observed between the LA and OA groups with respect to preterm delivery, the birth weight, the Apgar scores, the operative time, or intra-abdominal abscess formation after surgery. The results from this meta-analysis also showed that compared with OA, LA might be associated with a lower risk of wound infection and a shorter LOS.According to the current SAGES guidelines, LA may be performed safely in pregnant patients who have a suspicion of appendicitis . The preThe findings from the current updated meta-analysis showed that fetal loss was significantly higher in pregnant women who underwent LA compared with those who underwent OA. However, the sensitivity analysis revealed that this finding was greatly influenced by the study undertaken by McGory et al.  that hadThe effect of pneumoperitoneum is a major consideration in relation to laparoscopic surgery during pregnancy. Increasing the intra-abdominal pressure to induce pneumoperitoneum can reduce the venous return and cardiac output , resultiAlthough the precise cause of preterm delivery after surgery during pregnancy remains unclear, the findings from previous studies have indicated that preterm delivery was associated with uterine irritability during the operations , 45. WhiInterestingly, the pooled results from this updated meta-analysis that were related to the primary outcomes, namely, fetal loss and preterm delivery, showed opposing trends in the two groups. These results could be attributed to a single study even though the two outcomes may be caused by different factors, for example, pneumoperitoneum and uterine irritability during the operation. If fetal loss and preterm delivery are caused by pneumoperitoneum and uterine irritability, respectively, during the operation, it is very interesting that the pooled results showed opposing trends in the two groups.Postoperative complications are usually considered to evaluate the safety of surgical procedures. The results from this meta-analysis showed that the risk of wound infection was significantly lower in the LA group compared with that in the OA group. In contrast, the findings from a previous meta-analysis suggested that there was no significant difference between LA and OA during pregnancy with respect to wound infection . HoweverThe results from the present meta-analysis revealed that the mean LOS was significantly shorter in the LA group compared with that in the OA group, a finding that concurs with those reported from previous meta-analyses that used data from pregnant women and the general population , 46; henWe updated the meta-analysis that was conducted during a previous related study. This meta-analysis evaluated the effects of LA and OA on pregnancy and surgical outcomes. One of the study’s strengths was related to the sensitivity and meta-regression analyses of the pregnancy outcomes that were performed to corroborate the findings from the meta-analysis. Another strength of this study was the rigorous literature search. Despite these strengths, there are some limitations to the present study. First, there were no randomized controlled trials among the studies analyzed. Second, the studies analyzed tended to involve small study populations. Third, the outcome assessments differed and different definitions for each variable existed among the studies.In summary, the sensitivity analysis of the studies that comprised this meta-analysis showed that one study by McGory et al.  had a di


Objective: Verbal fluency tests (VFTs) are widely used in clinical practice and research to assess executive functions and are highly sensitive to frontal lobe lesions. However, using VFTs in different cultures and languages needs further considerations. The aim of this study was to provide a Persian (Farsi) version of verbal fluency with preliminary normative data. Method: In the first phase, 50 healthy native Persian-speaking individuals completed 1-minute VFT for all 32 letters in Persian to find letters with highest frequency. In the second phase, 100 healthy participants  were recruited into 5 age groups that were matched by gender and education. Participants were instructed to do 1-minute VFT for the 3 selected letters (phonemic VFT) and 3 categories  (semantic VFT). For data analysis, one-way ANOVA was performed. Results: In the first phase, 3 letters  had the highest frequency in word production (12 in average) and had been chosen for Persian phonemic VFT. Participants were assessed with the 3 selected letters  and 3 semantic categories  with 1-minute time limitation for each test. The results showed that education was significantly (p<0.01) associated with the performance in the phonemic but not semantic scores, while age was not correlated with either of the tests. No significant effect of gender was observed.Conclusion: Based on our results, we recommend Persian letters Pe, Meem, and Kaff that have the highest frequency in word production among others to be used for neuropsychological assessments and future studies in the Persian language. This is the same logic behind selecting F, A, and S in the English version. Although the norms obtained in this study are preliminary, these results can be useful in clinical evaluation with considerations about age and educational levels. Moreover, the findings of this study can be used as an initial step for more comprehensive normative studies. Verbal fluency tests (VFTs) have an important role in neuropsychological and clinical research and are included in almost every neuropsychological evaluation , 2. The VFTs measure retrieval speed of words under specific rules in a limited time. There are 2 types of verbal fluency tasks: (1) semantic or categorical; (2) phonemic or letter fluency. In semantic VFTs, participants are required to say as many words as they can from a specific category. The most common category is “animals”, but other categories such as “fruits and vegetables”, “things found in the kitchen”, “things to wear”, “things found in a supermarket”, and “vehicles” have also been used in some studies. In phonemic VFTs, participants are required to produce as many words as possible beginning with a specific letter. There are usually 3 letters in phonemic fluency tasks as there was in the original English version. F, A, and S are the most commonly used letters in English5, and other letter combinations, including C, F, L, P, R, and W, are also used , 10. TheThe necessity of language consideration for semantic fluency may be less obvious. Yet, many studies showed significant differences between speakers of different languages in semantic fluency in different categories that were not explained either by age and/or education or by vocabulary size , 11. ForIn an attempt to produce normative data for bilingual adolescent, Malek et al. 2013) developed a VFT for 302 bilingual (Turkish-Persian) adolescents -18 in Ta developeThus, if we want to use any language and/or culture bounded neuropsychological test out of the context they have been developed in, it is important to take necessary developmental steps to modify the tasks to suit the target culture and language and provide adequate norms considering demographic factors.Most of the published studies support the effects of demographic factors, mainly age, education, and gender, on verbal fluency tasks in both phonemic and semantic versions. Considering different studies with different age ranges, a nonlinear effect of age is apparent in both phonemic and semantic fluency. In phonemic fluency tasks, performance improves during childhood, and the scores increase from early childhood to adulthood. The increase peaks at about age 30 to 39 and shows a mild decline in older ages  and 34. Published reports for the influence of education levels on both phonemic and semantic fluency tasks are even more consistent. Higher level of education is associated with better performance on both tasks  and 33. The other demographic characteristic, which should be considered in the VFTs, is gender. Many studies found little or no evidence for gender effects on phonemic and semantic scores  and 34. Today, Persian is spoken primarily in Iran and Afghanistan, but it is also spoken by significant populations in other countries around the Persian Gulf  as well as by large communities in the USA. According to Penn language center , PersianIn this study, we aimed to developa Persian version of verbal fluency tasks and provide preliminary norms for it. To reach the mentioned goal, the study comprised of 2 phases: (1) Development phase; (2) Standardization phase.The sample for the first phase consisted of 47 healthy participants . Participants were recruited based on the following criteria:Age 18-60 years Being Iranian and Persian native speaker as their first mother tongue Having no medical disease that interferes with the assessments and no psychiatric disorder in both axis I and II of DSM-IV-TR or any other neurological or speech problems based on a structured clinical interviewIn the second phase, 100 healthy individuals  were classified into 5 age groups , and 2 groups according to educational level  in a balanced way in 20 clusters (gender*age*education level) with 5 participants each. The inclusion criteria were the same as the first phase. All participants were given informed consent forms, which included study conditions and benefits. Also, they were compensated for their time in the study with small gifts.Materials and Procedure In the first phase of this study, after a thorough instruction based on a written text and doing a sample trial to familiarize the participants with the test, they were asked to name as many words as they could, starting with specific letters in 60 seconds. Participants performed the task for all 32 letters in the Persian alphabet plus a specific vowel “آ” (aka Aa). The order of presenting letters was randomized to control the effects of learning and tiredness over different letters. Participants were instructed that proper nouns and multiple words using the same stem with a different suffix  were not acceptable. Also, the same verb conjugated in different tenses were not acceptable and counted as one. When a questionable response was provided, clarifications were made by the instructor.Time was measured by a stopwatch and participants’ answers were recorded. Final score for each letter was calculated after omitting errors and duplicates. Instructions were as follow: “I want you to say as many Persian words as possible that begin with a certain letter. You can say any word except for names of people and other proper nouns, such as Sarahor Tehran. Also, you should use different words rather than the same word with a different ending. For example, different forms of a verb or single and plural forms of a noun are not acceptable.” According to the results of the first phase, 3 letters  were selected for the Persian version of phonemic fluency test. These letters were selected because they had the highest frequency in world production among others , followiThe same procedure of administration and scoring were used for these 3 selected letters in the second phase to get a preliminary normative data among 100 participants. To measure semantic fluency in the second phase, 3 categories, “animals”, “fruits”, and “things found in the supermarket”, were presented to participants in random order. Instructions were as follow: “I want you to say as many Persian words as possible that belong to a certain category. You should use different words rather than the same word with a different ending. For example, single and plural forms of a noun are not acceptable.” For comprising the data, we used one-way ANOVA to compare age-related and educational factors for producing words.1 Sara2 TehranThe mean frequency of words produced in 60 seconds during the first phase of the study is shown in In the second phase, the total correct answers for the phonemic test with 3 letters and semantic test with 3 categories were calculated for statistical analysis. The mean score on the phonemic and semantic fluency for the whole sample was 37.30±9.58 and 62.44±11.89. Mean scores and standard deviations of phonemic and semantic fluency are presented according to different age groups, education level, and gender in The normality of sample was verified using the Kolmogorov-Smirnov test (K-S). One-way ANOVA results showed that phonemic fluency is significantly influenced by education  = 11.46, p =.001), but not by age groups  = 0.73, p = 0.593) or gender  = 0.66, p = 0.419). No interaction was observed between any of the variables tested by three-way ANOVA . No variP= 0.438; P = .000). In addition, all measures of phonemic fluency performance were positively correlated with education level; moreover, a weak but significant positive correlation was observed between animal fluency and education   was highly correlated (= 0.032) .The main objective of this study was to develop a verbal fluency test for a Persian speaking population and obtain preliminary normative data for healthy Iranian adults. The only existing norms for verbal fluency in Persian have used a restricted age range i.e., -18 and eThe current findings showed that education level is the only correlate of performance on phonemic fluency tests among basic demographic measures, as seen in other studies as well , 36. MeaEducation level was positively correlated with the phonemic score and its components, and a weak but significant relationship was observed between education level and animals’ category, but not other categories. The results indicated that the lowest education level was 12 years and the highest was 16 years. Education is still an important factor in generating words with specific letters. This finding is in contrast with a previous study )Evidence from a variety of sources shows that verbal fluency measures are relatively insensitive to gender , 35, 37.According to the results of this study, there was no significant difference between age groups in their verbal fluency. These results conform to some of the previous studies that reported non-significant age effects , 36. HowTwo major factors should be kept in mind when using the current norms. First, the norms are applicable only when these letters and categories are used. A degree of variety may exist among the number of words generated for different letters or categories. Second, although the formal language in Iran is Persian (Farsi), Iranian population consists of monolinguals and bilinguals, and in many ethnic groups in different regions of Iran, and the first language might not be the formal language. This difference may affect performance in verbal fluency for these groups and should be considered when conducting the tests in such regions and also in rural regions. One of the limitations of this study was the relative low number of participants from five different age groups with two different educational levels. Also, we did not assess other cognitive functions or IQ of the participants. Furthermore, all participants were selected from healthy population and we assumed that they had normal cognitive function.Taken together, to use a culture and language bound test such as verbal fluency in speakers of different languages, these tests must be modified and standardized in the target language under thorough examination. In this study, we selected letters in phonemic fluency task to suit our plan and provide primary normative data for healthy adults for the first time. We recommend Persian letters Pe, Meem, and Kaff that have the highest frequency in word production among others to be used for neuropsychological assessments and future studies in the Persian language. This is the same logic behind selecting F, A, and S in the English version. Likewise, our results showed the effect of education on phonemic tasks and revealed no age or sex effect on performance within this age range. The given data are expected to assist neuropsychologists and clinicians in evaluating adult Iranian patients and may help more confident clinical diagnoses.


Several genome‐wide association studies already explored the associations between 6q25.1 rs2046210 polymorphism and breast cancer (BC), but the results of these studies were not consistent. Thus, we conducted a meta‐analysis of relevant studies to better analyze the effects of rs2046210 polymorphism on individual susceptibility to BC.PubMed, Web of Science, and Embase were searched for eligible studies. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated.p < 0.0001, OR = 0.78, 95% CI 0.73–0.83), AA versus GG + GA , GA versus GG + AA , and G versus A . Further subgroup analyses yielded similar positive results in both Asians and Caucasians.Totally 21 studies with 261,703 subjects were analyzed. A significant association with BC was observed for the rs2046210 polymorphism in GG versus GA +AA (dominant comparison, In summary, our findings suggested that the rs2046210 polymorphism may serve as a potential genetic biomarker of BC in both Asians and Caucasians. It is the second most common cancer globally, and the most frequent cancer among females  conducted by zheng et al found that the rs2046210 G/A polymorphism at 6q25.1 was significantly associated with an elevated susceptibility to BC in both Chinese and Europeans  checklist  case–control study about rs2046210 polymorphism and BC; (b) providing sufficient data for calculating odds ratios (ORs) and 95% confidence intervals (CIs); (c) full text in English available. Studies were excluded if one of the following conditions was fulfilled: (a) not related to rs2046210 polymorphism and BC; (b) pedigree studies; (c) case reports or case series. In the case of duplicate reports by the same authors, we only included the most recent study.2.2p value) of Hardy–Weinberg equilibrium (HWE) was also calculated.We extracted the following information from eligible studies: (a) name of the first author; (b) year of publication; (c) country and ethnicity of participants; (d) sample size; and (e) the genotypic distribution of rs2046210 polymorphism in cases and controls. The probability value  to evaluate the quality of eligible studies Stang, . The NOSTwo reviewers conducted data extraction and quality assessment independently. When necessary, we wrote to the corresponding authors for extra information. Any disagreement between two reviewers was solved by discussion until a consensus was reached.2.3p value of 0.05 or less was defined as statistically significant. Between‐study heterogeneities were evaluated by I2 statistic. Random‐effect models (REMs) would be used for analyses if I2 was >50%. Otherwise, analyses would be conducted with fixed‐effect models (FEMs). Subgroup analyses were subsequently carried out by ethnicity. Stabilities of synthetic results were tested in sensitivity analyses. Publication biases were assessed by funnel plots.In the current study, we performed statistical analyses by using Review Manager Version 5.3.3. We calculated ORs and 95% CIs to estimate potential associations between rs2046210 polymorphism and BC in dominant (GG vs. GA + AA), recessive (AA vs. GG + GA), additive (GA vs. GG + AA) and allele (G vs. A) models, and a 33.1We found 171 articles by using our searching strategy. After excluding irrelevant and duplicate articles, 29 articles were retrieved for further evaluation. Another eight articles were subsequently excluded after reading the full text. Ultimately, a total of 21 eligible studies involving 131,785 cases and 129,918 controls were enrolled for analyses , AA versus GG +GA , GA versus GG + AA , and G versus A . Further subgroup analyses yielded similar positive results in both Asians and Caucasians (see Table Totally 261,703 subjects were analyzed. A significant association with BC was observed for the rs2046210 polymorphism in GG versus GA + AA (dominant comparison, 3.3We conducted sensitivity analyses by eliminating one individual study each time. The significant associations detected in pooled analyses remained unchanged in all comparisons, which suggested that our findings were statistically stable.3.4We used funnel plots to evaluate potential publication biases. The shape of funnel plots was symmetry for every comparison, which indicated that severe publication biases were unlikely.4To the best of our knowledge, this is so far the most comprehensive meta‐analysis about rs2046210 polymorphism and BC. The pooled analyses revealed that the rs2046210 polymorphism was significantly associated with BC in both Asians and Caucasians. The stabilities of synthetic results were evaluated by sensitivity analyses, and no alterations of results were observed in any comparisons, which suggested that our findings were statistically stable. As for evaluation of heterogeneities, significant heterogeneities were detected in every comparison of overall analyses, and thus all analyses were performed with REMs. But in further subgroup analyses, a reduction tendency of heterogeneity was found for Asians, which suggested that differences in ethnicity could partially explain observed heterogeneities between studies.ESR1, the encoder of ERα. Since it was evident the binding of estrogen and ERα could result in increased proliferation of normal and cancerous breast epithelial cells (Ali & Coombes, ESR1 and consequentially influence individual susceptibility to BC. Secondly, the etiology of BC is extremely complex, and as a consequence, to better elucidate potential roles of genetic variations in BC, we strongly recommend future studies to conduct haplotype analyses and investigate potential gene–gene interactions.There are several points that worth noting about this meta‐analysis. Firstly, the 6q25.1 rs2046210 polymorphism is located within 1‐Mb upstream of Some limitations of this meta‐analysis should also be noted when interpreting our findings. First, our pooled analyses were based on unadjusted estimations due to lack of raw data, but we have to admit that failure to perform further adjusted analyses may impact the reliability of our findings (Xie, Shi, & Liu, 5In summary, our meta‐analysis suggested that the rs2046210 polymorphism may serve as a potential genetic biomarker of BC in both Asians and Caucasians. However, further well‐designed studies are still needed to confirm our findings.6For this type of study formal consent is not required.This article does not contain any studies with human participants or animals performed by any of the authors.The authors declare that they have no conflict of interest.Tie‐feng Jin and Zhen‐feng Zhou conceived of the study, participated in its design. Tie‐feng Jin and Wen‐ting Zhang conducted the systematic literature review. Tie‐feng Jin and Wen‐ting Zhang performed data analyses. Tie‐feng Jin and Zhen‐feng Zhou drafted the manuscript. All authors have read and approved the final manuscript.


Studies evaluating the association between abdominal obesity and asthma yielded conflict results. Whether abdominal obesity is positively associated with asthma remains unclear.To quantitatively determine the association between abdominal obesity and asthma.Databases including PubMed, Web of Science, China National Knowledge Infrastructure, China Biology Medicine disc, Chinese Scientific and Technological Journal Database and Wanfang Data were searched up to February 2018 to collect all relevant studies. Reference lists of related articles were also checked. After study selection and data extraction, meta-analysis was conducted to calculate the pooled odds ratio (OR) and corresponding 95% confidence interval (CI). Subgroup analyses by study design and age groups of participants were further performed. Publication bias was assessed via Begg’s rank correlation and Egger’s linear regression methods.2 = 10.7%) or publication bias  was found. Subgroup analyses by study design and age groups of participants obtained consistently positive results across subgroups. Moreover, our meta-analysis observed similar results when considering this association separately in males and females . In addition, the association between abdominal overweight and asthma was further explored in this meta-analysis and the pooled OR and 95% CI was 1.13 , indicating that there is a dose–response relationship between abdominal weight status and asthma.A total of 13 studies were included in the final meta-analysis, including 2 case–control studies, 6 cohort studies, and 5 cross-sectional studies. Our meta-analysis observed a positive association between abdominal obesity and asthma . No evidence of heterogeneity (IOur meta-analysis shows a positive association between abdominal obesity and asthma. Moreover, this association is similar in males and females. In addition, our meta-analysis indicates that there is a dose–response relationship between abdominal weight status and asthma. Therefore, addressing abdominal obesity issue is of great importance. More studies are needed in the future to clarify the association between abdominal obesity and asthma. As one of the most common chronic respiratory diseases, asthma affects about 358 million people worldwide . It is pObesity, as is known to all, is also a worldwide public health problem. In 2015, approximately 107.7 million children and 603.7 million adults worldwide were obese, accounting for about 30% of the world’s population , 9. A toA growing number of studies have showed a significant association between obesity and asthma –13. MoreIn spite of this, studies on the association between abdominal obesity and asthma are lacking. The findings of the existing studies are inconsistent. A case–control study  demonstrTherefore, we conducted this meta-analysis to quantitatively determine the association between abdominal obesity and asthma. Different from previous meta-analyses published in this field –17, our Databases including PubMed, Web of Science, China National Knowledge Infrastructure, China Biology Medicine disc, Chinese Scientific and Technological Journal Database and Wanfang Data were searched up to February 2018 to collect all relevant studies by two researchers independently. The search strategy was as follows:  AND . In addition, reference lists of related original and review articles were also checked to find additional studies that may meet the eligibility criteria.Eligible studies had to meet the following criteria: (1) were original epidemiological studies (of any design); (2) related to the association between abdominal obesity and asthma; (3) clearly stated the definition and measurement of abdominal obesity; (4) reported odds ratios (OR) or risk ratios (RR) and their 95% confidence intervals (CI), or provided sufficient data from which these measures could be calculated; (5) published in English or Chinese. Studies were excluded if they: (1) were animal studies; (2) were editorials, comments, or literature reviews; (3) reported overlapped data; (4) contained incomplete data which was still unavailable after contact with the author.The titles and abstracts of the retrieved studies were screened according to the above criteria by two researchers independently. The full texts of the potentially relevant studies were then obtained and strictly assessed. Disagreements were resolved by discussion with a third researcher.Data extraction was conducted by two researchers separately. Discrepancies were resolved by discussion with a third researcher. The following data were extracted from each eligible study: first author, country, year of publication, year of study, study design, study population, sample size, definition of abdominal obesity, anthropometric measures of abdominal obesity, adjusted OR or RR and their corresponding 95% CI, and confounders that were controlled in the study. If no effect estimate was provided in a given study, OR or RR and 95% CI were calculated from the raw data presented in the study. The authors were contacted for further information when necessary.For cohort studies and case–control studies, the quality were assessed by the Newcastle–Ottawa Scale (NOS). The NOS evaluates a study based on three major aspects: the selection of the study groups, the comparability of the groups, and the ascertainment of the exposure or outcome of interest . For cro2 statistic was used to test for the heterogeneity across studies. If no evidence of heterogeneity was presented (I2 < 50%), a fixed effect model was used. Otherwise, a random effect model was adopted. Sensitivity analysis was conducted by excluding one study at a time to determine whether the results of this meta-analysis were robust. Subgroup analyses based on study design and age groups of participants were further performed. Publication bias was assessed when there were no fewer than 10 studies, via Begg’s rank correlation and Egger’s linear regression methods. Probability value P < 0.05 was considered statistically significant. All statistical analyses were performed using STATA software .Where the identification of abdominal obesity adopted consistent anthropometric measures across studies and the final results were presented in a similar fashion, meta-analysis was performed to calculate the pooled OR and associated 95% CI. Moreover, sex-specific association was further explored in this meta-analysis. When a study reported data for males and females separately, a fixed effect model was used to calculate the pooled OR, which was then used as the measure for the overall population of the study. The IA total of 114 studies were identified, of which 103 studies were identified through database searching, 11 studies were identified by reference lists checking. After removal of duplicates, the titles and abstracts of the remaining 92 studies were screened for eligibility. The full texts of the 23 potentially related studies were then strictly assessed. Finally, 13 studies were included in the meta-analysis  or publication bias  was found. After sequentially excluding each included study in sensitivity analysis, the results remained largely unchanged, indicating that the meta-analysis was stable.Only data of WC were analyzed in this meta-analysis, since measures other than WC were rarely used in the 13 included studies. Our meta-analysis observed a positive association between abdominal obesity and asthma . Of the 2 studies, one study [It is noted that 2 , 33 of tne study  reportedA total of 7 studies reported data separately for males and females. Therefore, the sex-specific association between abdominal obesity and asthma was further explored in this meta-analysis.2 = 0.0%). No significant change was found in sensitivity analysis. For males, similar results were observed  and asthma –17, our Therefore, on the basis of our findings, addressing abdominal obesity issue is of great importance. There is growing evidence showing that in addition to the increased risk of developing asthma, abdominal obesity is also associated with lower lung function, more severe asthma symptoms, as well as more poorly controlled asthma , 36, 37.2 adrenergic receptor and tumour necrosis factor ɑ have been found to be strongly associated with both obesity and asthma [Although the clear mechanism underlying the association between abdominal obesity and asthma remains largely unknown, hypotheses can be suggested. On the one hand, the accumulation of abdominal adipose tissue mechanically restricts the diaphragm and limits lung expansion, resulting in reduced functional residual capacity and tidal volume, which further leads to the conversion of airway smooth muscle from rapidly cycling actin-myosin cross bridges to slowly cycling latch bridges –44. The d asthma , 45. Somd asthma , 45. Besd asthma , 44.There are some limitations in this meta-analysis. First, due to the lack of study on the association between abdominal obesity and asthma, we included all available relevant original epidemiological studies (of any design) in this meta-analysis. As a result, three types of study, cross-sectional study, cohort study, as well as case–control study, were included in our meta-analysis. However, different types of study may reflect different indexes of asthma. For example, of the studies included in our meta-analysis, 5 cross-sectional studies reflected the association between abdominal obesity and the current prevalence of asthma, while the 6 cohort studies reported the association between abdominal obesity and the incidence of asthma. Therefore, the results of our meta-analysis cannot be simply interpreted as abdominal obesity is associated with the prevalence or with the incidence of asthma. It may be more appropriate to consider our results as the association between abdominal obesity and the overall risk of asthma. However, according to our subgroup analysis by study design, all three subgroups obtained statistically significant results, indicating that abdominal obesity may be associated with both the prevalence and the incidence of asthma. More studies are needed in the future to clarify the association between abdominal obesity and asthma.Second, although our meta-analysis focused on the association between abdominal obesity (measured by WC) and asthma, the results of our meta-analysis were similar to those of previous meta-analyses that focused on general obesity (measured by BMI). Therefore, whether abdominal obesity add further information to the currently recognized relationship between general obesity and asthma is in doubt. However, based on the current evidence, we cannot resolve this doubt in this meta-analysis. But among the 13 included studies, 2 studies , 33 repoThird, in the majority of the included studies (10 of 13 studies), the definition of asthma was based on self-report, which was subject to reporting bias and misclassification. However, self-reported asthma is widely used in epidemiological studies and has been evaluated as a reliable and valid measure for asthma –51. MoreFourth, only data of WC were adopted in our meta-analysis as there were few studies have provided data on other abdominal obesity measures. However, using multiple measures of abdominal obesity to comprehensively evaluate the association between abdominal obesity and asthma is also desirable. Therefore, future studies can use other measures of abdominal obesity to better explore this relationship.Despite the limitations mentioned above, there are also some strengths in this meta-analysis. First, different from previous meta-analyses that focused on the association between general obesity (measured by BMI) and asthma –17, our This meta-analysis shows a positive association between abdominal obesity and asthma. Moreover, this association is similar in males and females. In addition, this meta-analysis suggests that there is a dose–response relationship between abdominal weight status and asthma. Therefore, addressing abdominal obesity issue is of great importance. More studies are needed in the future to clarify the association between abdominal obesity and asthma. Studies adopting other measures of abdominal obesity are also desirable.


TNF-α) gene are emerging as key determinants of gastric diseases. The TNF-α-238G/A single-nucleotide polymorphism (SNP) is the most extensively studied. However, this association is inconsistent amongst different populations. We therefore conducted an updated meta-analysis to obtain a more precise estimate of the association of TNF-α-238G/A polymorphism with gastric cancer (GC) risk. A comprehensive search of PubMed, Embase, Chinese (CNKI and WanFang) databases was performed to identify relevant studies through 5 May 2018. Odds ratio (OR) and 95% confidence interval (CI) were used to assess the strength of the association. Fourteen studies were included in our meta-analysis involving 2999 cases and 4685 controls. There was no significant association between TNF-α-238G/A polymorphism and GC risk in the overall populations. In the subgroup analysis, we found that TNF-α-238G/A polymorphism was associated with the increased risk of GC amongst Asians, especially in Chinese, but not in Caucasians. Subgroup analysis by genotyping methods revealed increased risk for other methods. In conclusion, our present meta-analysis shows that TNF-α-238G/A polymorphism is associated with the risk of GC in East Asian individuals.Polymorphisms in the tumor necrosis factor α ( Helicobacter pylori (H. pylori), which progresses through a multi-step process, developing from gastritis to gastric atrophy, intestinal metaplasia, dysplasia, and finally to carcinoma [H. pylori infection does not always result in a high incidence of GC. Only 1–2% amongst those infected will develop stomach cancer [Gastric cancer (GC) is the fourth most common malignancy worldwide , with a arcinoma . Howeverh cancer , suggestTNF-α gene, mainly in the 5′-promoter regions. Several SNPs  in TNF gene have been widely reported, in which the positions -308 and -238 have been most frequently evaluated for association with GC [Polymorphisms of inflammation-related genes have provided evidence that host genetic factors play a pivotal role in the pathogenesis of GC ,6. TNF-α with GC .The association between GC and TNF-α-308 G/A polymorphism has been confined by two meta-analyses ,13. ThisAdditionally, Cen and Wu  and WangIn summary, only the -238G/A polymorphism exists with conflicting results, although some published meta. In addition, considering the important role of TNF-α in gastric carcinogenesis and some limitations in the previous two meta-analyses, we performed an updated analysis on all eligible case–control studies to estimate the GC risk associated with -238 G/A polymorphism . To our knowledge, this is the most updated meta-analysis conducted to date with respect to the association between TNF-α-238 G/A polymorphism and GC risk.TNF-α gene and GC risk were retrieved. Studies that were included in our meta-analysis had to meet all of the following criteria: (i) evaluation of TNF-α gene -238 G/A polymorphism and GC risk; (ii) case–control design; (iii) availability of genotype frequency; (iv) availability of full text; and (v) genotype distributions of control consistent with HWE. Meanwhile, the following exclusion criteria were also used: (i) no control population, (ii) no available genotype frequency, (iii) HWE of controls were <0.05, and (iv) studies have not been published; for studies with overlapping or repeating data, the most recent or complete studies with the largest number of cases and controls were included and others were excluded.A literature search of the PubMed, Embase, Web of Science, Google Scholar, WanFang, and CNKI database (updated on 11 June 2018) was conducted using combinations of the following keywords ‘polymorphism’ or ‘variant’ or ’mutation’, ‘gastric’ or ‘stomach’, ‘cancer’ or ‘carcinoma’ and ‘TNF’ or ‘tumor necrosis factor alpha’. There was no language restriction. All studies that evaluated the associations between polymorphisms of The following data were collected from each study: first author’s last name, year of publication, race of origin, sample size (cases/controls), each frequency number of genotype both case and control samples, study design  and population-based (PB)), HWE of controls and genotype method.Risk ratios (ORs) with 95% confidence intervals (CIs) were used to measure the strength of the association between TNF-α gene -238 G/A and GC based on the genotype frequencies in cases and controls. We analyzed this relationship between TNF-α gene -238 G/A polymorphism and GC risk using three different genetic models: allelic contrast (A compared with G), heterozygote comparison (AG compared with GG), and dominant genetic model (AA+AG compared with GG). Different ethnic descents were categorized as Caucasian and Asian. Subgroup analysis was stratified by a source of control and genotype methods.Q-test. The statistical significance of the summary OR was determined with the Z-test. The heterogeneity amongst the studies was checked using the chi-square based Q statistic and considered statistically significant at P<0.10. When P for heterogeneity test (Ph)>0.10, the pooled OR of each study was calculated using the fixed-effects model ; otherwise, the random-effects model (the DerSimonian and Laird method) was used [P<0.05 considered as statistically significant [2 test in controls using the Pearson chi-square test, again with P<0.05 considered as significant [http://biostat.mc.vanderbilt.edu/wiki/Main/PowerSampleSize#Windows). All statistical tests for this meta-analysis were performed with Stata software  [Heterogeneity assumption was evaluated with a chi-square-based was used ,30. Funnnificant . The depnificant . The powion, TX) .Genotyping of TNF-α gene -238 G/A was conducted using different techniques in different studies: GeneChip, Sequence, Taqman, Snapshot, PCR-restriction fragment length polymorphism (RFLP), and PCR-based denaturing HPLC (DHPLC).http://gepia.cancer-pku.cn/) [To further explore the association between TNF-α expression and GC, we used the bioinformatics web: GEPIA (pku.cn/) , which pA total of 145 articles were retrieved based on our selection strategy from the PubMed, Embase, Web of Science, Google Scholar, WanFang, and CNKI databases. Amongst them, 10 duplicated articles were excluded; 61 articles were excluded after reviewing the title or abstract. Seventy-four articles were then downloaded for review of the full text. Amongst them, 55 articles were excluded, 1 was not a case–control study, 1 did not provide detailed genotyping information, 30 investigated other polymorphisms within the TNF gene, 2 articles did not conform to inclusion criteria, and 3 articles contained the same or partly same individuals with other studies, the larger or the latest one was chosen . After tWe also checked the minor-allele frequency (MAF), the frequency of the most common mutant allele, of the TNF-α gene -238 G/A polymorphism amongst the five main population groups reported in the 1000 Genomes Browser: East Asian (0.0308), European (0.0636), African (0.0378), American (0.0821), and South Asian (0.1053) . The MAFPheterogeneity<0.01, P=0.384, Pheterogeneity<0.01, P=0.423), and the dominant allele model   .Pheterogeneity=0.024, P=0.039, Pheterogeneity=0.041, P=0.040; AA+AG compared with GG: OR = 1.40, 95% CI = 1.02–1.93, Pheterogeneity=0.024, P=0.040). In the analysis stratified by country, similarly significant associations were detected between TNF-α gene 308 G/A polymorphism and GC risk in the Chinese population   .Pheterogeneity=0.114, P=0.001; AG compared with GG: OR = 1.58, 95% CI = 1.00–2.52, Pheterogeneity=0.072, P=0.051; AA+AG compared with GG: OR = 1.60, 95% CI = 1.01–2.53, Pheterogeneity=0.073, P=0.045, In the analysis stratified by genotype methods subgroup, a similar magnitude of association was observed between TNF-α gene -238 G/A polymorphism and GC risk in the group for others .Sensitivity analysis was performed to assess the influence of each individual study on the pooled OR by sequential removal of individual studies. The results suggested that no individual study affected the overall OR significantly . This suIn silico results indicated that the expression of TNF in GC tissue was higher than in normal stomach (TPM: transcripts per kilobase million = 0.76 compared with 0.23 respectively, P<0.01, P=0.34, pHR = 0.35, P=0.98, pHR = 0.99, GC remains a high-incidence malignance and the incidence greatly varies between countries. The majority of cases are registered in developing countries, with half of them reported in Eastern Asia. In addition, GC incidence is twice as high in men as in women. This suggests that race, environmental, hormonal, or genetic factors may affect GC risk . FurtherFour previous meta-analyses ,48 focusin silico analysis, the expression of TNF-α was higher in GC tissues, which was similar to carcinogenic genes and was consistent with previous publications. Maolake et al. [In our e et al.  suggestee et al.  showed te et al.  reportede et al. , indicatMeta-analyses have been recognized as an effective method to summarize and review previously published quantitative research to answer a wide variety of clinical questions . HoweverIn summary, our meta-analysis showed that TNF-α gene -238 G/A polymorphism was associated with significantly increased GC risk in East Asian populations, especially Chinese. Furthermore, well-designed and larger studies, dealing specifically with gene–gene and gene–environment interactions, are warranted.


This meta-analytic study explored the relationship between the risk of type 2 diabetes mellitus (T2DM) and bisphenol A concentrations.The Embase and Medline (PubMed) databases were searched, using relevant keywords, for studies published between 1980 and 2018. A total of 16 studies, twelve cross-sectional, two case-control and one prospective, were included in the meta-analysis. The odds ratio (OR) and its 95% confidence interval (CI) were determined across the sixteen studies. The OR and its 95% CI of diabetes associated with bisphenol A were estimated using both fixed-effects and random-effects models.A total of 41,320 subjects were included. Fourteen of the sixteen studies included in the analysis provided measurements of urine bisphenol A levels and two study provided serum bisphenol A levels. Bisphenol A concentrations in human bio-specimens showed positive associations with T2DM risk . A sensitivity analysis indicated that urine bisphenol A concentrations were positively associated with T2DM risk .This meta-analysis indicated that Bisphenol A exposure is positively associated with T2DM risk in humans.The online version of this article (10.1186/s12902-018-0310-y) contains supplementary material, which is available to authorized users. Type 2 diabetes mellitus (T2DM) is a metabolic disease that presents with symptoms of insulin resistance and lack of insulin . The gloBisphenol A (BPA) is a role for endocrine disrupting chemicals (EDCs) and especially used in epoxy resin and polycarbonate plastic products such as food packaging, drink containers, and dental sealants –7. Once Competitive with 17- beta estradiol (E2), BPA is a type of endocrine disrupting chemical (EDCs) that disrupts estrogenic response by binding to estrogen receptors. BPA binds to androgen receptors and thyroid receptors. Unfortunately, humans are exposed to BPA through the daily exposure to BPA containing products such as canned food, plastic products, dental sealants, and household dust .In recent studies, research findings suggest that low levels of BPA can cause significant health problems. A number of scientists hypothesized that adverse health effects might be associated with high urinary BPA concentrations. Epidemiological studies have been carried out to evaluate the possible association between BPA exposure and the risk of T2DM, but the results were not consistent –24.In this study, a meta-analysis focusing on the association between BPA concentrations (measured in urine or serum) and the risk of T2DM was performed. In addition, subgroup analyses were performed according to the sample type (urine or serum) and the study design.Figure The keywords used in the Embase and Medline (PubMed) database searches were: Bisphenol A, BPA, 4, 4 isopropylidenediphenol or Bisphenol A bis (2 hydroxypropyl) ether dimethacrylate and Noninsulin dependent diabetes mellitus or Type 2 diabetes or Diabetes Mellitus, Type 2 or Diabetes Mellitus, Noninsulin-Dependent or Diabetes Mellitus, Ketosis-Resistant or Diabetes Mellitus, Fasting blood sugar or Fasting plasma glucose or Blood glucose, HbA1c or Glycosylated hemoglobin or Hemoglobin A1cor Glycated Hemoglobin A or Hemoglobin A, Glycated. A total of 420 articles were found: 246 were from Embase and 174 were from Medline (PubMed). First, 139 duplicated articles were removed., After, an initial review, 148 studies were excluded; 85 studies were not human research such as animal and invitro experiment, 43 studies had irrelevant exposures or outcomes and 20 studies were reviews or meta-analyses papers. Next, 133 studies were selected for full-text article review. From these studies, 117 studies, including 44 studies were not human research, 38 with irrelevant exposures or outcomes, 10 studies were reviews or meta-analyses, 22 were letter or book or comment papers, and 3 had not find full text from the same database were excluded. Finally, a total of 16 articles were included in this meta-analysis , and type of study  as possible sources of heterogeneity. A Begg’s Funnel Plot and an Egger’s Regression Test were conducted to minimize publication bias and asymmetry of the studies. When publication bias exists, the Begg’s Funnel Plot is asymmetric, or the Egger’s Test P-value < 0.05 [Odds ratios (OR) and 95% confidence intervals (95% CI) were obtained from the selected articles using the standard guidelines for meta-analysis . Fixed-ee < 0.05 .To adjust for the cross-study differences between the BPA concentration units and the range of measured values, a dose-response meta-analysis (DRMA) was implemented. The dose-response meta-analyses (DRMA) was implemented by using the STATA GLST command  on a samStatistical analyses were performed using STATA version 13.0 software .The 420 studies were searched using a systematic search strategy, referring to the PRISMA flow chart that describes the selection process of the meta-analysis . After tTable BPA exposure was positively associated with the risk of T2DM .Previous studies have identified that the association between urinary BPA levels and T2DM may be biologically feasible. For example, BPA, an estrogen agonist that acts as an endocrine hormone disruptor, has been shown to be involved in several mechanisms of diabetes development including glucose homeostasis, obesity, insulin resistance, beta-cell dysfunction, inflammation, and oxidative stress . BPA binPrevious meta-analyses, have not implemented the dose-response analytic method used in this study to determine the relationship between BPA exposure and the risk of T2DM. A significant dose-response relationship was found between urinary BPA concentrations (mg/dL) and T2DM risk. In addition, subgroup analysis was performed according to the type of sample (urine or serum), and the type of study . Moreover, quality assessment methods were implemented to remove any irrelevant studies and to improve the validity of the meta-analysis. The current study considered only diabetes as a primary outcome variable as was the case with the final sixteen articles used for the meta-analysis. Despite the fact that diabetes mellitus is an important risk factor for cardiovascular disease, our study will focus specifically on T2DM, and as a result will be more focused and statistically significant than previous studies.There were several limitations to the conduct and analysis in this study that must be considered. First, because of the limited number of cohort studies investigating the relationship between BPA exposure and T2DM risk that have been conducted, this meta-analysis included only thirteen cross-sectional, two case-control studies and one prospective studies. The inclusion of additional studies are required to validate and confirm these results. Second, this meta-analysis included fourteen studies which used spot urinary BPA concentrations and two study that used serum BPA concentrations as a surrogate marker of BPA exposure. It is unclear whether spot urinary BPA concentrations could accurately reflect the long term exposure level of BPA in individuals. Spot urinary BPA concentrations are the most commonly used method to assess BPA exposure levels because of it is short half-life and the convenience of the measurement method , 20. AltIn conclusion, this meta-analysis demonstrated that BPA concentrations measured in urine or serum is positively associated with T2DM risk. Furthermore, prospective cohort studies, including carefully collected data about the dietary sources of BPA exposure and potential confounding, will help clarify the role of BPA in the pathogenesis of diabetes.Additional file 1:PRISMA 2009 Checklist. Preferred report items for systematic review and meta-analysis were identified through a checklist. (DOC 64 kb)Additional file 2:Figure S1. Funnel plot with egger. Figure S2. GLST. These are additional sub-analysis results. (PPTX 44 kb)


ADIPOQ) polymorphisms are associated with the risk of polycystic ovary syndrome (PCOS) remain controversial. Therefore, we performed this study to better explore correlations between ADIPOQ polymorphisms and PCOS risk.Whether adiponectin  and 95% confidence intervals (CIs) were calculated.p = 0.02, OR = 0.77, 95%CI 0.62–0.95; allele model: p = 0.001, OR = 1.15, 95%CI 1.06–1.26). Further subgroup analyses according to ethnicity of participants revealed that rs1501299 and rs2241766 polymorphisms were both significantly correlated with PCOS risk in Caucasians. In addition, rs1501299 polymorphism was also significantly correlated with PCOS risk in East Asians.Eighteen studies were enrolled for analyses. Pooled overall analyses showed that rs1501299 polymorphism was significantly associated with PCOS risk (recessive model: Our findings indicated that rs1501299 and rs2241766 polymorphisms might serve as genetic biomarkers of PCOS in certain ethnicities. Polycystic ovary syndrome (PCOS), featured by oligomenorrhea, polycystic ovaries, anovulatory infertility, hyperandrogenism, insulin resistance or hyperinsulinemia, and an elevated risk of multiple metabolic diseases, is an extremely common reproductive endocrine disorder, with an estimated prevalence of approximately 5–10% in women of childbearing age . AlthougAdiponectin (ADIPOQ), a multifunctional adipocytokine that is primarily secreted by adipocytes, plays a pivotal role in regulating energy and material metabolism . PreviouADIPOQ gene located on chromosome 3q27 [ADIPOQ polymorphisms, rs1501299 and rs2241766, were correlated with altered serum concentration of adiponectin [ADIPOQ polymorphisms and PCOS risk, but the results of these studies were controversial [ADIPOQ polymorphisms in PCOS.Adiponectin is encoded by the ome 3q27 . It was ponectin , 15. As oversial –33. TherThis meta-analysis was adhered to the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guideline . PotentiADIPOQ polymorphisms and PCOS risk; (2) provide genotypic and/or allelic frequency of investigated ADIPOQ polymorphisms in cases and controls; (3) full text in English or Chinese available. Studies were excluded if one of the following criteria was fulfilled: (1) not relevant to ADIPOQ polymorphisms and PCOS; (2) case reports or case series; (3) abstracts, reviews, comments, letters and conference presentations. For duplicate publications, we only included the study with the largest sample size for analyses.To test the research hypothesis of this meta-analysis, included studies must meet all the following criteria: (1) case-control study on correlations between ADIPOQ polymorphisms in cases and controls. Additionally, the probability value  of Hardy-Weinberg equilibrium (HWE) was also calculated. When necessary, we wrote to the corresponding authors for raw data. We used the Newcastle-Ottawa scale (NOS) to assess the quality of eligible studies [The following data were extracted from included studies: (1) the name of the first author; (2) publication time; (3) country and ethnicity; (4) sample size; and (5) genotypic distributions of  studies . This scADIPOQ polymorphisms and PCOS risk in all possible genetic models, and p values ≤0.05 were considered to be statistically significant. Between-study heterogeneities were evaluated with I2 statistic. If I2 was greater than 50 %, random-effect models (REMs) would be used to pool the data. Otherwise, fixed-effect models (FEMs) would be employed for synthetic analyses. Subgroup analyses by ethnicity were subsequently performed. Sensitivity analyses were conducted to examine the stability of synthetic results. Funnel plots were used to evaluate possible publication bias.All statistical analyses were conducted with Review Manager Version 5.3.3 . Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to estimate strength of associations between We found 331 potential relevant articles. Among these articles, a total of 18 eligible studies were finally included for synthetic analyses  polymorphism in overall analyses. Further subgroup analyses according to ethnicity of participants revealed that rs1501299 and rs2241766 polymorphisms were both significantly correlated with PCOS risk in Caucasians. In addition, rs1501299 polymorphism was also significantly correlated with PCOS risk in East Asians (see Table To investigate potential correlations between We performed sensitivity analyses by excluding studies that deviated from HWE. No alterations of results were detected in sensitivity analyses, which suggested that our findings were statistically reliable.Publication biases were evaluated with funnel plots. We did not find obvious asymmetry of funnel plots in any comparisons, which indicated that our findings were unlikely to be impacted by severe publication biases.ADIPOQ polymorphisms and PCOS risk. Our overall and subgroup analyses demonstrated that rs1501299 and rs2241766 polymorphisms were both significantly correlated with PCOS risk in Caucasians. Moreover, rs1501299 polymorphism was also significantly correlated with PCOS risk in East Asians.To the best of our knowledge, this is so far the most comprehensive meta-analysis on correlations between There are several points that need to be addressed about this meta-analysis. Firstly, previous experimental studies showed that mutant alleles of investigated polymorphisms were correlated with decreased adiponectin generation, which may partially explain our positive findings , 15. SecADIPOQ polymorphisms and PCOS risk may also be modified by gene-gene and gene-environmental interactions. However, most eligible studies ignore these potential interactions, which impeded us to perform relevant analyses accordingly [As with all meta-analysis, this study certainly has some limitations. First, our results were derived from unadjusted analyses due to lack of raw data, and lack of further adjusted analyses for potential confounding factors may impact the reliability of our findings . Second,ordingly . To sum In conclusion, our meta-analysis suggested that rs1501299 and rs2241766 polymorphisms might serve as genetic biomarkers of PCOS in certain ethnicities. However, further well-designed studies are still warranted to confirm our findings.