The relationship between hypothyroidism and primary open angle glaucoma (POAG) has attracted intense interest recently, but the reported results have been controversial. This meta-analysis was carried out to determine the association between hypothyroidism and POAG.The literature was identified from three databases . The meta-analyses were performed using random-effects models, with results reported as adjusted odds ratios (ORs) with 95% confidence intervals (CI 95%).P < 0.001; I2 = 83.2%). Sub-group analysis revealed that the cohort studies and case–control studies showed a significant association between hypothyroidism and POAG, which was not observed in cross-sectional studies. There was no significant publication bias in this study.A total of 11 studies meeting the inclusion criteria were included in the final meta-analysis. The pooled OR based on 11 risk estimates showed a statistically significant increased risk of POAG prevalence among individuals with hypothyroidism . Substantial heterogeneity among these studies was detected (The findings of this meta-analysis indicate that individuals with hypothyroidism have an increased risk of developing POAG. Glaucoma, which is the major cause of irreversible blindness worldwide, is a coSmith et al.[Hypothyroidism, which affects nearly 10% of the general population and occurs more frequently among those of advanced age, is a prevalent endocrine disease characteth et al. also fouth et al.,13,14.However, there have been contradictory reports in the literature about whether hypothyroidism is independently associated with an increased risk of developing POAG. Several previously published studies have reported significant associations. Kim et al. and GirkSince this question is unlikely to be answered through an individual epidemiological study, the main aim of this study was to determine the potential relationship between hypothyroidism and POAG by using a systematic review and meta-analysis.Three databases were used to conduct a computerized search of terms from their inception to February 01, 2017: Web of Science, Embase, and PubMed. The following terms were used in the search: hypothyroidism, thyroid disease, thyroid disorder, thyropathy, glaucoma, ocular hypertension, intraocular pressure, and intraocular hypertension. No language restrictions were applied. A hand search for other potential articles was carried out by checking the reference lists of the original paper.The inclusion criteria for research included in this meta-analysis were as follows: (1) cross-sectional, case–control, or cohort design, (2) evaluated the relationship between hypothyroidism and POAG, and (3) reported relative risk (RR) or odds ratios (ORs) estimates with 95% confidence intervals (CIs) . The exclusion criteria were the following: (1) studies involving angle-closure glaucoma or secondary glaucoma, (2) the data provided could not be used to calculate the RRs or ORs, or (3) case reports, abstracts, reviews, and reports with incomplete data. If different publications from the same study subjects were available, the most recent one was included.Two reviewers extracted the following information independently: first author, year of publication, study design, country of the subjects studied, methods used to determine hypothyroidism, definition of glaucoma, age of the study participants, sample size, outcomes measured with 95% CIs, and the adjusted variables. Two reviewers independently assessed study quality using the tool described by Sanderson et al. The vari2 index. For the Q statistic, P < 0.05 indicated statistically significant heterogeneity. For the I2 statistic, if I2 > 50%, heterogeneity was said to exist[P < 0.05 was considered statistically significant in the test for overall effect.This meta-analysis was conducted using Stata 12.0 . Fully adjusted summary estimates were used to assess the relationship between hypothyroidism and POAG; moreover, the adjusted summary estimates and their corresponding 95% CIs were pooled using the random-effects model. A homogeneity test across studies was performed using a Cochran’s Q test and the I to exist. Subgrou to exist,22. P < The initial database search yielded 1,165 potential entries, of which 392 were duplicate publications and thus removed. Of the 773 remaining publications that qualified for title and abstract review, 756 were excluded because of apparent irrelevance, e.g., non-human studies, reviews, or topics other than hypothyroidism. Consequently, 17 studies remained for further assessment and a full-text review. Among them, three studies were excluded because they did not provide sufficient data to calculate the OR,24,25. AThe characteristics of the included epidemiological studies are shown in Pheterogeneity < 0.001; I2 = 83.2%). Considering the significant heterogeneity found among the studies, a set of subgroup analyses were performed based on study design, geographical area, methods for determining hypothyroidism, publication year, and the number of adjusted variables. In terms of stratified study design, the results showed that there was a significant positive association between hypothyroidism and POAG with a case–control design  and nested case–control/cohort design , but not in studies with a cross-sectional design . Next, subgroup analyses were performed according to geographical area. Hypothyroidism was observed to be significantly associated with POAG in studies performed in North America  and Asia . However, this association was not found in studies performed in Europe . The pooled OR was not consistent in studies where hypothyroidism status was ascertained via self-reports as well as by medical records. An association was observed between hypothyroidism and POAG in methods for determining hypothyroidism via medical records, but not in methods for determining hypothyroidism via self-reports. The impact of confounding factors on the estimate of OR was also considered. When studies adjusted for > 3 confounder factors, the summary OR showed that hypothyroidism might increase the risk of POAG; conversely, when studies adjusted for ≤ 3 confounder factors, a negative association was observed between hypothyroidism and POAG. Another subgroup analysis based on publication year was conducted, with the results showing that the 1993–2010 publication period did not significantly affect this association. In contrast, a positive association was detected between hypothyroidism and POAG in the 2010–2015 publication period. However, significant heterogeneity was still observed in some of the subgroups. Detailed information about the subgroup analysis is presented in Based on combined results calculated using the random-effect model of the 11 included studies, this meta-analysis showed that hypothyroidism might increase the risk of POAG. As shown in 2 = 12.0%; P = 0.332). After deleting Shim’s[2 = 34.7%; P = 0.133)  . The pubion bias .Although several risk factors for the development of POAG, such as diabetes, increasIn order to confirm this potential association, several epidemiological studies have tried but failed to reach a consistent conclusion. Thus, it remains controversial whether or not hypothyroidism increases the risk of POAG. Meta-analysis is a powerful statistical method that allows data on the same topic to be synthesized with greater power to obtain more accurate risk estimations. Meta-analysis was therefore conducted to reach a definite conclusion about the relationship between hypothyroidism and POAG. This study, which was based on 11 previous studies containing a total sample of 381,695 subjects, demonstrated a statistically significant association between hypothyroidism and POAG. Individuals with hypothyroidism were 1.64 times more likely to have POAG than those without hypothyroidism. Subgroup analysis showed that unlike European individuals, North American and Asian individuals had an increased risk of developing POAG. The small sample sizes and resulting insufficient statistical power of the two included studies from Europe might explain this negative association. In addition, it should be noted that in subgroup analyses of study design, cross-sectional design did not show an association; this might be explained by the significant selection bias the cross-sectional design was subject to. To achieve a reliable and convincing result, a series of analyses were carried out. Sensitivity analyses were conducted, by which the influence of a single study on the pooled effect was examined by removing one study at a time. The results showed that this analysis did not significantly change the global estimation, which supports the reliability of this study. Additionally, publication bias analysis showed a low probability of publication bias in the pooled results, further demonstrating the robustness of the present meta-analysis.Stein et al.[There are some possible mechanisms that could support the notion that hypothyroidism affects susceptibility to and the progression of POAG. One such mechanism is the deposition of mucopolysaccharides and hyaluronic acid in the trabecular meshwork caused by hypothyroidism, which in turn causes an obstruction in outflow that would elevate the IOP. Anotherin et al. demonstrThis meta-analysis showed that hypothyroidism was observed to significantly increase the risk of POAG, although the substantial heterogeneity between effect size estimates limits the ability to draw firm conclusions. This heterogeneity may reflect different eligibility criteria, study populations, definitions of POAG and hypothyroidism, and/or levels of adjustment for potential confounders. To investigate the sources of heterogeneity, stratified analyses according to study design, geographical area, methods for determining hypothyroidism, publication year, and the number of adjusted variables were performed. However, only slight changes were found, and a high level of heterogeneity remained in most of the subgroups. Then, in the sensitivity procedure, it was found that heterogeneity among the included studies could be attributed to Motsko’s and ShimThe major strength of this study was that it performed the first meta-analysis to assess the association between hypothyroidism and POAG via an extensive literature search that included as many relevant studies as possible to obtain a more precise conclusion. Another strength of the present study is that the enlarged sample size gave the meta-analysis enhanced statistical power to obtain a more reliable estimation of the association between hypothyroidism and POAG. Lastly, the studies included in this meta-analysis were conducted in different countries, making the results more generalizable.Several limitations of this study should be acknowledged. First, the diagnosis of hypothyroidism in several of the studies was ascertained based on self-reports, which could cause the misclassification of hypothyroidism patients as non-hypothyroidism subjects. This underestimation might attenuate the true correlation between hypothyroidism and POAG. Second, the sample sizes in some subgroups were too small and had insufficient statistical power to reach a positive association. Third, some of the included studies failed to adjust for common confounding variables known to be risk factors for POAG, which might have affected the pooled results. Fourth, the possibility of publication bias is a major problem in any meta-analysis, because statistically significant results are more likely to be published than those with null results. Nevertheless, we found no evidence of publication bias using Begger’s and Egger’s tests in this meta-analysis. Finally, significant heterogeneity existed among the studies. Thus, the findings of this meta-analysis should be interpreted with caution.In conclusion, the findings of this meta-analysis suggest that individuals with hypothyroidism have an increased risk of developing POAG. Given the limitations of this meta-analysis, this conclusion should be interpreted with caution. In the future, both experimental and epidemiological studies are needed to better understand the association between hypothyroidism and POAG.S1 Table(DOC)Click here for additional data file.


Serum gamma-glutamyltransferase (GGT) elevation likely contributes to cardiovascular (CV) mortality, however it has remained unknown whether a dose-response relationship exists between serum GGT and CV mortality.We searched the PubMed, EMBASE, and Cochrane library databases for prospective cohort studies published up to October 2, 2016. Summary hazard ratios (HRs) with their corresponding 95% confidence intervals (CIs) were calculated using a fixed effects model.Nine prospective studies, including 527,589 participants and more than 7,011 cases, were included in this meta-analysis. For the moderate, high, and highest levels of GGT, the pooled HRs of CV mortality were 1.11 (95% CI = 1.04–1.19), 1.29 (95% CI = 1.21–1.38) and 1.59 (95% CI = 1.47–1.72), respectively . Additionally, the HR per incremental increase of GGT by 10 U/L was 1.10 (95% CI = 1.08–1.11). Evidence of a positive relationship with nonlinear trend for GGT elevation with CV mortality in females was found (P = 0.04 for nonlinearity). However, a linear model was better fit to illustrate the GGT-CV mortality among males (P = 0.304 for nonlinearity).These findings indicate that serum GGT activity within the reference interval is positively associated with increased risk of CV mortality in a dose-response manner. Cardiovascular disease (CVD) is one of the leading causes of mortality in developed countries and responsible for an estimated 17.5 million annual deaths in the world . DespiteSerum gamma-glutamyltransferase (GGT), a commonly used marker of hepatobiliary disease and excess alcohol consumption ,5, is a Despite the mechanism being unknown, there are considerable prospective studies published reporting on the independent role of serum GGT in the pathogenesis and clinical evolution of cardiovascular diseases ,16 and CWe followed the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guideline  for perfThe included studies in the meta-analysis had to meet the following criteria: 1) have a prospective cohort design ; 2) subjects enrolled in the study at baseline were free of any pre-existing diseases, especially cardiovascular diseases in the general population; 3) the exposure of interest was serum GGT concentration and the corresponding categories were more than 2 levels; 4) the outcome of interest was CV mortality; 5) Reported adjusted HRs or relative risks (RRs) with 95% CIs at least three quantitative GGT categories, or provided the number of cases and total participants or person-years for each category of serum GGT levels; 6) the reported HRs or RRs had been adjusted at least for age and gender; 7) the duration of follow-up was more than five years.One investigator (J.N.W) extracted the data from the eligible studies, another investigator (D.D.Z) validated the data for accuracy independently with a standardized form as follows: first author’s name, publication year, country of the participants, name of study or source of participants, baseline survey period, follow-up time (year), age at recruitment (mean or range), gender , mean or median range of GGT levels in each category, the size of observational population and CV mortality cases for each GGT levels, values of HRs or RRs with their 95% CIs by quantile, and adjusted potential confounders. Additionally, if the articles presented the data separately by gender, we treated them as two independent studies and extracted the data separately. When diverse adjustments were provided, we extracted the HRs (95% CIs) with the most confounders adjusted. For studies with published data in people with and without diabetes separately, we just extracted the results in subjects without diabetes. The summary data of each identified work in our meta-analysis were presented in We conducted separate meta-analyses for different levels of GGT concentration categorized into four levels as in a meta-analysis of Coffee Consumption and Risk of Cardiovascular Disease . We calcTo identify whether there is a potential nonlinear relationship between GGT levels and CV mortality, a two-stage dose-response meta-analysis was performed. We used the generalized least-squares (GLST) calculation described by Orsini and colleagues  and multFurthermore, we performed subgroup analysis according to study location, gender, age, follow-up time, sample size, adjustment of diabetes mellitus, body mass index (BMI) and alcohol intake. To check whether the results could have been affected markedly by a single study, diverse sensitivity analyses were carried out by removing one study at a time and recalculating the pooled estimates for the remainder of the studies.2 statistic  [Heterogeneity among studies was quantitatively evaluated using the chi-square-based Q test  and Igeneity) . A randogeneity) . Potentigeneity) , Begg’s geneity)  and Eggegeneity)  with P vA total of nine independent prospective articles –24,26 coCompared to the subjects with the lowest GGT categories in fourteen studies, the pooled HRs (95% CI) for CV mortality incidence were 1.11  for the moderate, 1.29  for the high and 1.59  for the highest categories of GGT levels. The P values for heterogeneity of the Q test in the above analyses were 0.64, 0.079 and 0.409, respectively, moderate heterogeneity was only upon analyzing the high versus the lowest categories of GGT .nonlinearity < 0.001. see 2 = 55.8%, P = 0.012)  . However= 0.012) . In cont= 0.012) , indicat= 0.012) . Nonethe= 0.012) . A weighinteraction = 0.029. In addition, there was significant difference between subgroups when stratified by gender and age (P < 0.05 for inter-subgroup comparison). The results of sensitivity analyses are present in To examine the stability of the primary results, we carried out subgroup analyses across the 11 studies ,22,23,26There was no evidence of publication bias for any association as revealed by Begg’s test and Egger’s test .We performed a meta-analysis of the moderate, the high, and the highest levels of serum GGT compared to the lowest level of serum GGT and dose-response relationship between GGT and the incidence of CV mortality. There was evidence of a nonlinear positive dose-response relationship between GGT and CV mortality after pooling seven articles including 515,446 participants and 6,339 cases of CV mortality. Notably, a flat association was observed in the lowest range of GGT levels < 13U/L, thereafter the shape of dose-response curve was steeper with serum GGT levels 13 -< 52.5U/L. Furthermore, the increases in risk somewhat flattened at above 52.5U/L GGT. For the dose-response meta-analysis, increases of 10U/L of circulating GGT levels were associated with a 10% increase in the risk of CV mortality. Sensitivity analysis was conducted further and did not significantly alter the result. No evidence of heterogeneity or publication bias was detected across the 9 prospective articles. Compared with the lower categories, those of the moderate or highest GGT activity categories did have a significantly increased risk of CV mortality in all participants combined and in subgroups. There is an incremental increase in CV mortality of approximately 9% for males and 12% for females per 10 U/L of GGT elevation, this association seems to be stronger for females than males. However, the current findings should be interpreted critically, considering that the GGT level of comparison groups was quite different.The results from the British Women’s Heart and Health Study indicated that GGT activity did not have a causal role in the incidence of coronary heart disease (CHD), independent of alcohol intake . HoweverSeveral researchers have also reported that the association between serum GGT activity and CV mortality appears to be stronger in younger versus older subjects ,18,20,22Similar to previous meta-analyses ,28, our 2 = 55.8%, P = 0.012), therefore, the results should be interpreted with caution.Based on the RR or HR values for every dose category in each study included in the dose-response analysis, the association between serum GGT activity and the risk of CV mortality was significant in 1 study  and was For subgroup analyses of the dose-response meta-analysis, the incidence of CV mortality increased with rises in GGT concentration in all subgroups with moderate to substantial heterogeneity, which did not alter much in the subgroup analyses. There are differences in serum GGT levels between males and females in our included studies. Therefore, within the subgroup analysis we examined gender as possible sources of heterogeneity. Females were more sensitive to develop CV mortality on the same GGT elevation degree than males (P < 0.05), revealing that gender might be confounding the GGT-CV mortality association. When stratified by age, our subgroup analysis revealed a stronger relationship on GGT-CV mortality in the older (P < 0.05). However, several researchers have reported that the association between serum GGT activity and CV mortality appears to be stronger in younger versus older subjects ,18,20,22heterogeneity = 0.029 for both comparisons), with a stronger association among studies that adjusted alcohol consumption, so it seems plausible that the relationship between GGT and CV mortality risk varies by alcohol consumption. In addition, GGT has been demonstrated to be positively associated with several cardiovascular disease risk factors, such as BMI [We also examined region, number of participants, and length of follow-up as possible sources of heterogeneity, nevertheless, heterogeneity between studies did not alter much in the subgroup analyses. There is substantial evidence that alcohol consumption has been reported to be U-curve or J-curve association with CAD –65. As Gh as BMI  and diabh as BMI , whereasCompared with the previous meta-analyses , our stuLimitations of this meta-analysis should be mentioned. First, all included studies measured GGT level on a single occasion at baseline. Considering the within-subject variation of GGT level and measurement error, our results might be subject to regression dilution bias , indicatAltogether, our meta-analysis contributes to the literature by exploring the nonlinear dose-response relationship between GGT and the risk of CV mortality. These results indicated that GGT in normal range might be an individualized predictor in screening incidents of CV events. However, the potential mechanisms are still limited. Further research is necessary to further clarify the mechanisms and examine the potential modifiers of the GGT-CV mortality relationship.S1 AppendixThe PRISMA Checklist of this meta-analysis.(DOC)Click here for additional data file.S2 Appendix(PDF)Click here for additional data file.S1 Table(PDF)Click here for additional data file.S2 Table(PDF)Click here for additional data file.


MALAT-1) is one kind of long non-coding RNAs (lncRNAs) that has been recognized as a hallmark of the onset and development of several carcinomas. This study seek to meta-analyze the overall diagnostic efficacy of elevated MALAT-1 expression profile for human cancers. Studies on the diagnostic performance of MALAT-1 in cancers were retrieved by searching the online databases. The combined effect sizes were summarized using a bivariate meta-analysis model. Impacts of publication bias on the pooled effect sizes were assessed using “Duval and Tweedie nonparametric trim and fill method”. Sensitivity analysis and meta-regression test were applied to deeply trace the heterogeneity sources among eligible studies. A total of 14 studies with 1342 cancer cases were included. The combined effect sizes showed that MALAT-1 expression profiling conferred an estimated sensitivity of 0.69 (95% CI: 0.62–0.75) , specificity of 0.85 (95% CI: 0.79–0.90)  and AUC (area under curve) of 0.83 in distinguishing cancer patients from noncancerous contrasts. Moreover, stratified analysis depending on cancer type manifested that elevated MALAT-1 harbored a promising efficacy in the diagnosis of pulmonary tumors (AUC = 0.90), digestive system tumors (AUC = 0.84), gynecologic cancers (AUC = 0.84) and nasopharyngeal carcinoma (AUC = 0.84), particularly in confirming the subtype of squamous carcinoma (AUC = 0.91) and non-small cell lung carcinoma (AUC = 0.88) in lung cancer. Other analyses based on test matrix and ethnicity also presented robust results. Collectively, elevated MALAT-1 could be developed as an auxiliary molecular marker to aid in cancer diagnosis.Metastasis-associated lung adenocarcinoma transcript 1 ( Cancer is now becoming a global burden that has increased massive economic and social pressures around the world. According to the global cancer statistics data in 2012, nearly 14.1 million new cancer cases and 8.2 million cancer deaths were reported worldwide [MALAT-1) is one kind of nuclear lncRNAs that implicates in a spectrum of biological processes in vertebrate cells [MALAT-1 is overexpressed in multiple types of human malignancies, such as the pulmonary cancers, digestive system tumors and genitourinary cancers, and so forth [MALAT-1 could be rated as a promising biomarker to aid in cancer diagnosis and prognosis [MALAT-1 in predicting metastasis and/or survival in cancers has been documented by some meta-analysis studies [MALAT-1 for carcinomas. In order to elucidate the global diagnostic efficacy of MALAT-1 in identification of cancers and provide valid evidences, we conducted this meta-analysis according to standard statistical methods. Our data may help to better understand the clinical value of elevated MALAT-1 in the diagnosis of various cancers.In human beings, the majority of the genome is transcribed, yet only 2% accounts for protein-coding exons . Evidencte cells , 6. MALAso forth –20. Notarognosis –22. The  studies , 22. To MALAT-1 in cancer were included in the statistical analysis.The search of relevant articles was undertaken following the procedures of the PRISMA diagram Figure . BrieflyMALAT-1 were determined based on the approach of quantitative real-time polymerase chain reaction (qRT-PCR), and that GAPDH [β-actin [HPRT190 [PSAKIT [All essential data were collected from 14 studies, comprising a combined population of 1342 cancer patients and 1189 noncancerous controls. All cancer patients had a definite diagnosis based on histopathological examination. All samples were obtained by surgical operation or biopsy prior to other therapies, and the specimen sources were from plasma , 15, 18,at GAPDH , 18–20, [β-actin , 15, 17, [PSAKIT  were utiProportions of risks on bias and applicability by the QUADAS-2 checklist are shown in Figure MALAT-1 expression harvested a combined sensitivity of 0.69 (95%CI: 0.62–0.75) and specificity of 0.85 (95%CI: 0.79–0.90) in discriminating cancers from noncancerous controls, corresponding to an AUC value of 0.83 . Additionally, in the subgroup studies by type of carcinoma, test matrix and ethnicity, heterogeneity also appeared in the analyses of pulmonary tumor , adenocarcinoma (AdCa) , squamous carcinoma (SqCC) , gynecologic cancer  as well as in serum-  and Asian-based  analyses , especially in confirming the subtypes of SqCC (AUC = 0.91) and non-small-cell lung cancer (NSCLC) (AUC = 0.88). Of note, the pooled specificities in pulmonary tumor, AdCa and SqCC were shown to be 0.92 (95% CI: 0.85–0.96), 0.91 (95% CI: 0.76–0.97) and 0.94 (95% CI: 0.85–0.98), respectively. Moreover, the diagnostic efficacy of MALAT-1 in other cancers, including digestive system tumor, gynecologic cancer and nasopharyngeal carcinoma also revealed robust results (Table MALAT-1 testing presented an AUC of 0.88 better than that of serum- (AUC = 0.85), tissue- (AUC = 0.77) and urine-based (AUC = 0.65) analyses, indicating that plasma might be a suitable test matrix for the analysis of MALAT-1. Additionally, testing depending on ethnicity displayed that Asian- and Caucasian-based MALAT-1 analysis retained equal AUC values (0.82), whereas Caucasian-based analysis harbored a superior specificity of 0.93 (95% CI: 0.90–0.96) and DOR of 19.23 (95% CI: 10.92–33.88) , but an increase of heterogeneity in specificity (I2 from 87.95% to 88.22%). Besides that, the pooled NLR dropped from 0.37 to 0.35, and DOR elevated form 12.56 to 12.71. Further univariate meta-regression test was undertaken depending on the covariates of cancer type, specimen source, sample size, reference gene and study quality [P = 0.0012) is more like to be a cause of study heterogeneity.Sensitivity analyses of the overall pooled effect size showed that 1 individual study by He et al.  was esti quality . As summP < 0.05) displayed obvious publication bias among studies. In consequence, the “Duval and Tweedie nonparametric trim and fill method” was employed to elucidate the possible effects of bias on the pooled analysis [P < 0.001) before adjustment versus that of 0.166 (P < 0.001) after adjustment, suggesting that the pooled effect was slightly altered before and after adjustments. Evaluation of the publication bias in the subgroup studies was performed as well and no clear bias was detected among stratified analyses  and NSCLC (AUC of 0.88). Importantly, MALAT-1 testing showed promising specificities higher than 0.90 in confirming lung cancer , as well as the AdCa and SqCC subtype. Analysis of the efficacy in other types manifested that MALAT-1 testing harvested an AUC of 0.84, specificity of 0.88 and DOR of 11.33 in identifying nasopharyngeal carcinoma, suggesting that MALAT-1 might be developed as a promising biomarker for nasopharyngeal carcinoma as well. Besides that, MALAT-1 showed equal AUC values of 0.84 in both of the digestive system and gynecologic tumors, but a higher combined specificity of 0.83 was observed in the latter.In the present study, heterogeneity seems to be existed among the overall pooled analyses mainly due to the included different types of cancers. In consequence, we further conducted subgroup studies depending on the type of carcinoma, test matrix and ethnicity. Our results revealed that elevated MALAT-1 testing presented a better AUC value than the serum-, tissue- and urine-based analyses, indicating that plasma might be a suitable matrix for the analysis of MALAT-1 expression in cancers. On the other hand, our subgroup analysis by region showed that Asian- and Caucasian-based MALAT-1 testing sustained equal AUC values (0.82), whereas the latter harbored a superior specificity (0.93) and DOR (19.23). Notwithstanding, the Caucasian-based analysis were only from 1 study  and  and/or (“diagnosis” or “area under the curve” or “AUC” or “sensitivity” or “specificity” or “ROC” or “Receiver operation characteristic curve”). We also manually searched the attached references in articles to increase search sensitivity.The international databases included PubMed/Medline, Evidence-Based Medicine Reviews, EBSCO, and BioMed Central were searched for the retrieval of eligible articles in English, and that CNKI, Wanfang and Weipu databases were retrieved for obtainment of studies published in Chinese. Date of publication was set up to May 1st, 2017. The search approaches with Medical Subject Heading terms or free-text words were utilized as: MALAT-1 in cancer(s); (2) expression of MALAT-1 was assessed by quantitative real-time polymerase chain reaction (qRT-PCR) or microarray analysis or other approaches; (3) studies had at least a disease group and a control group, with a sample size larger than 20; (4) the estimated sensitivity, specificity or AUC were available; and (5) the full-text was published in English or Chinese. Studies did not match the following criteria were excluded: (1) the control types were undefined or the sample sizes were smaller than 20; and (2) studies identified as review articles, basic research, animal studies, comments, letters or conference abstracts.Studies were firstly included if they in accordance with the following criteria: (1) studies evaluated the diagnostic performance(s) of www.quadas.org) [The bias among eligible studies was evaluated in duplicates by two group authors, using the evidence-based Quality Assessment for Studies of Diagnostic Accuracy II (QUADAS-2) checklist ( das.org) . This evThe basic information of articles were extracted and collected in twice by two trained person, including author's name, article publication date, study population, cancer type, sample size and type, control size and type, test method, reference gene, sensitivity, specificity, AUC (area under curve), cut-off value, etc. Any disagreements during data extraction will be solved by group consensus .Chi2 (Chi-squared), and I2 (I-squared) tests as well as the L’Abbe and Galbraith plot analysis. Either P < 0.05 for the Chi2 test or I2 > 50% for the I2 test were both considered as pooled effects with significant inconsistency. The pooled effect sizes included sensitivity, specificity, PLR, NLR, and DOR were generated using either a random-effect model or a fixed-effect model depending on study heterogeneity. The underlying sources of heterogeneity were further traced by sensitivity analysis and univariate meta-regression test [P < 0.05. The possible effects of publication bias on the overall pooled accuracy were assessed using “Duval and Tweedie nonparametric trim and fill method” [We conducted and reported this systematic meta-analysis in terms of the guidelines of the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRASMA) statement . All staion test . Study b method” .


The etiological cause of ovarian cancer is poorly understood. It has been theorized that bacterial or viral infection as well as pelvic inflammatory disease could play a role in ovarian carcinogenesis.To review the literature on studies examining the association between ovarian cancer and bacterial or viral infection or pelvic inflammatory disease.Neisseria gonorrhoeae”, “Mycoplasma genitalium”, “Papillomaviridae”, or “pelvic inflammatory disease”. Corresponding searches were performed in EMBASE, and Web of Science. The literature search identified 935 articles of which 40 were eligible for inclusion in this review.Database search through MEDLINE, applying the medical subject headings: “Ovarian neoplasms”, AND “Chlamydia infections”, “Mycoplasma genitalium in a large proportion of ovarian cancer cases. The remaining studies found no association. Human papillomavirus detection rates varied from 0 to 67% and were generally higher in the Asian studies than in studies from Western countries. Cytomegalovirus was the only other virus to be detected and was found in 50% of cases in a case-control study. The association between ovarian cancer and pelvic inflammatory disease was examined in seven epidemiological studies, two of which, reported a statistically significant association.Seven studies examined the association between bacterial infection and ovarian cancer. A single study found a significant association between chlamydial infection and ovarian cancer, while another study identified Data indicate a potential association between pelvic inflammatory disease and ovarian cancer. An association between ovarian cancer and high-risk human papillomavirus genotypes may exist in Asia, whereas an association in Western countries seems unlikely due to the low reported prevalence. Potential carcinogenic bacteria were found, but results were inconsistent, and further research is warranted.The online version of this article (doi:10.1186/s13027-017-0134-9) contains supplementary material, which is available to authorized users. Ovarian cancer (OC) is a major threat to female health, with a global prevalence of 239,000 cases in 2012 [Almost 90% of OC originates from epithelial cells, which have been thought to arise from the surface mesothelial lining of the ovaries. However, recent research has indicated that a proportion of serous epithelial OC could originate in precancerous lesions called “serous tubal intraepithelial carcinomas” (STICs) located in the fimbriated end of the fallopian tubes . This isThe female internal genitalia and the peritoneal cavity are accessible to outside pathogens through the genital tract. The mesothelial lining of the peritoneal cavity is identical in both sexes. However, primary serous peritoneal cancer, which may represent a continuum with serous OC , occurs To review the existing literature on OC and infectious agents, we performed a systematic literature search by means of a database search for English language articles published in the period from 1980 to 2016. Using the Pubmed research tool, the keyword: “ovarian neoplasms” was by means of the Boolean logical “AND”, combined with the following keywords: “Chlamydia infections” (MeSH), “Neisseria gonorrhoeae” (MeSH), “Mycoplasma genitalium” (MeSH), “Papillomaviridae” (MeSH), “pelvic inflammatory disease” (MeSH). Correspondent searches were performed in EMBASE and Web of Science. The initial search identified 935 articles. After removal of 130 duplicates, 805 studies remained for further analysis. Contents were outside the scope of the present review in 740 articles that were excluded. The remaining 65 articles were subjected to full-text screening. Original studies, examining the association between infectious agents and ovarian cancer by use of tissue-based or serologic methods were included. Epidemiological studies examining the association between pelvic inflammatory disease and ovarian cancer were also included. Case-reports and reviews were excluded, leaving a total of 40 articles to be included in the review. More detailed information on the search strategy is given in the “Additional file Table HPV infection is a prerequisite for cervical cancer development and is implicated in a range of other cancer forms such as vulvar, penile, anal, and oropharyngeal cancers. To date, The International Agency for Research on Cancer has classified 12 HPV subtypes as human carcinogens: HPV16/18/31/33/35/39/45/51/52/56/58/59 . The carHPV is the most studied infectious agent in relation to OC, and the available studies are summarised below.Six case-control studies, with 11–50 cases of OC, were identified that examined the relation between HPV and OC –17. The case-series and cross-sectional studies reporting on the prevalence of HPV in OC tissue were identified  database have been utilised to detect HPV oncogene expression in ovarian tumours. Using such data, Roos et al. concluded that six samples in 405 (1.5%) were positive for HPV18 oncogene expression . MoreoveCytomegalovirus (CMV) is under increasing focus as a carcinogenic virus. It is speculated that rather than initiating malignant transformation, CMV can enhance the growth and migration of tumour cells through a process called “oncomodulation” . For insEpstein-Barr virus (EBV) is present in more than 90% of the global adult population and causes latent infection in the host.  It exhibFinally, a single study has analysed tissue samples for the presence of the BK virus and John Cunningham virus . The oncChlamydia trachomatis (C. trachomatis), Mycoplasma genitalium , and Neisseria gonorrhoeae (N. gonorrhoeae) [In relation to OC, previous research has focused on the potential association between rrhoeae) , 57–61. C. trachomatis is a small, obligate intracellular bacterium that has developed several mechanisms to improve its own survival and replication within the host. It can upregulate the production of reactive oxygen species (ROS) through the Mitochondrial Nod-like Family Member NLRX1 [C. trachomatis can subvert cellular defences and use ROS to improve its own growth through activation of caspase-1 [C. trachomatis can impair the DNA damage response (DDR) through inhibition of DDR proteins pATM and 53BP1 [C. trachomatis can initiate proteasomal degradation of P53 through interaction with the phosphorylated ubiquitin ligase Murine Double Minute 2, leading to inhibition of apoptosis [C. trachomatis is a suitable candidate for ovarian carcinogenesis. However, the literature on the subject is scarce, and an association between C. trachomatis and OC has been found in only two studies.er NLRX1 . ROS proaspase-1 . The resnd 53BP1 . Despitend 53BP1 . Finallypoptosis . Combinepoptosis . TherefoC. trachomatis DNA by PCR techniques. C. trachomatis DNA was detected in 87.5% of the malignant tumours and 50% of the borderline ovarian tumours as compared to none of the benign ovarian tissue samples .One case-control study was based on fresh ovarian tissues from 39 women undergoing laparotomy . The tisC. trachomatis was performed. Patients with EOC were more likely to have high levels of C. trachomatis EB than the controls in adjusted analysis. In a subgroup analysis, the probability of having OC was insignificantly increased in patients with the highest levels of IgG EB antibodies compared to the group with the lowest levels (OR = 1.9 95% CI 0.9–3.9). However, these findings are controversial, since four other serologic studies with a total of 626 cases of OC found no association [A study by Ness et al. included 117 women with OC and 171 healthy controls . Serologociation –59, 61. ociation .N. gonorrhoeae is a small, gram-negative diplococcus that infects epithelial cells and causes cervicitis/urethritis and PID. It has been associated with prostate and bladder cancer [N. gonorrhoeae can cause DNA strand breaks, affect the cell cycle progression, and can decrease the levels of P53 within infected cells [r cancer , 67. Oneed cells .M. genitalium is under increasing focus for its role in PID. Like N. gonorrhoeae, its propensity for malignant transformation of host cells is unclear. However, one study revealed that persistent infection of benign human prostate cells with M. genitalium resulted in increased migration/invasion [invasion . Furtherinvasion .M. genitalium or N. gonorrhoeae and OC. Only one study reported an association, M. genitalium DNA being detected in 16 (59.3%) out of 27 samples by a combined PCR and ELISA technique. [M. genitalium or N. gonorrhoeae and OC [Very few studies have investigated the association between chnique.  However,e and OC , 57.Microorganisms that cannot directly induce cellular transformation may still play a role in OC oncogenesis due to the paradoxical effect of the host inflammatory response. It has long been recognised that the cells involved in innate and adaptive immune responses are recruited to the site of tumorigenesis. Here, they can release a multitude of tumour promoting inflammatory cytokines, chemokines, and ROS that can potentially facilitate tumour cell migration, metastasis, and angiogenesis . Cancer The reviewed studies were very heterogeneous in terms of study population, study design, and the analysis methods used.However, based on the reviewed studies, it is fair to conclude that high-risk HPV is unlikely to be associated with OC in Western countries. Interestingly, a higher prevalence was consistently reported in Asia and the Middle East. This finding is supported by previous reviews focusing on HPV and ovarian cancer , 81 and Most of the epidemiological studies found an association between OC and PID, but only two studies reported a statistically significant association , 75. It Interestingly, CMV was reported in 50% of cases in a single study, but this finding requires confirmation in future and larger studies. No other viral agent was detected.C. trachomatis and M. genitalium. The results of this review do not support an association between C. trachomatis and OC, the majority of studies reporting negative results. The association between M. genitalium and OC was investigated and found in only a single study and more research is needed.The only detected bacteria were Several factors may account for the conflicting results. First of all, there may be no connection between the investigated infectious agents and OC. Secondly, the heterogeneity of analysis methods, study populations, and study designs may play a part. However, other explanations must be discussed. Importantly, it is unclear to what extent signs of microbiological agents are present in a tumour that arises decades after infection. For some oncogenic viruses, like high-risk HPV, expression of viral oncogenes is obligate for the maintenance of the malignant phenotype . In thesFinally, other oncogenic agents act more indirectly by inducing chronic inflammation that may function as an initiator or promoter of carcinogenesis . ConsequC. trachomatis antibody titres 6 years after the diagnosis of PID [These examples highlight the fact that the abscense of microbiological agents in tumour tissue does not rule out their potential role in the transformation of host cells. These issues seem to strengthen the need for serologic studies. However, serologic methods are limited by the fact that antibody levels are likely to decline over time. In a study from Finland, 43% of women had declining s of PID . Moreoves of PID .These obstacles call for a revised strategy in the attempt to uncover the potential role of infectious agents in OC. It is fair to assume that signs of microbiological presence will be easier to detect when the interval between primary infection and investigation is short. Therefore, it seems more relevant to investigate precursor lesions for the presence of potential microbiological agents . Until rThe reviewed articles indicate a potential association between PID and OC. An association between OC and high-risk HPV genotypes may exist in Asia, whereas an association in Western countries seems unlikely due to the low reported prevalence. Two studies reported an association between bacterial agents and OC, but the majority of bacterial studies reported negative findings. However, the available literature was scarce and more studies are warranted.


I2=58.2%, P=0.026). The average Newcastle–Ottawa scale (NOS) score was 7.29, suggesting a high quality. There was no publication bias in the meta-analysis about dietary zinc intake on the risk of pancreatic cancer. Subgroup analyses showed that dietary zinc intake could reduce the risk of pancreatic cancer in case–control studies and among American populations. In conclusion, we found that highest category of dietary zinc intake can significantly reduce the risk of pancreatic cancer, especially among American populations.Previous reports have suggested a potential association on dietary zinc intake with the risk of pancreatic cancer. Since the associations between different studies were controversial, we therefore conducted a meta-analysis to reassess the relationship between dietary zinc intake and pancreatic cancer risk. A comprehensive search from the databases of PubMed, Embase, Web of Science, and Medline was performed until January 31, 2017. Relative risk (RR) with 95% confidence intervals (CI) derived by using random effect model was used. Sensitivity analysis and publication bias were conducted. Our meta-analysis was based on seven studies involving 1659 cases, including two prospective cohort studies and five case–control studies. The total RR of pancreatic cancer risk for the highest versus the lowest categories of dietary zinc intake was 0.798 (0.621–0.984), with its significant heterogeneity among studies ( Pancreatic cancer is one of the leading causes of cancer-related mortality worldwide –3. Of thThe pancreas is both an endocrine and an exocrine organ. Zinc is involved in a multitude of these processes within the pancreas including glucagon secretion, digestive enzyme activity, and insulin packaging, secretion, and signaling. As a result of this extensive physiological contribution, dysregulation of zinc metabolism may be associated with pancreatic cancer . A publiAn electronic search of PubMed, Embase, Web of Science, and Medline was performed until January 31, 2017. The keywords imputed are ‘zinc’ or ‘diet’ or ‘lifestyle’ or ‘Zn’ combined with ‘pancreatic cancer’ or ‘pancreatic carcinoma’ with no language or publication year restriction. The full text of relevant citations from all the results identified has been inspected and analyzed. Relative references in the main outcomes have also been searched and reviewed. The study selection process was performed following the PRISMA .Evaluating all the studies above that present quantitative estimates regarding the linkage between dietary zinc intake and pancreatic cancer, and those studies that meet the requirement were embraced in our research and then used for meta-analysis. We made the following strict criteria for our studies: (1) The study design was prospective cohort, case–control, or cross-sectional. (2) Human population studies instead of animals such as mice or rats. (3) The outcome of interest was pancreatic cancer. (4) The independent variable of interest was dietary zinc intake. (5) The risk estimates such as relative risk (RR) or odds ratio (OR) with 95% confidence intervals (CI) were reported or the numbers of case and control and the total numbers were reported. The studies could not satisfy such criteria were ruled out immediately.We established a standard data collective form to arrange the data of interest. The data extracted from the studies in use are referring such aspects: author name, year of publication, design of study, sex of population, age, number of cases and participants, value of RR or OR with 95% CI, and relative adjustments. A third reviewer was sought to make a common consensus on the abstracted data. The methodological quality of each study was assessed separately, using Newcastle–Ottawa scale (NOS)  to assesI2, which depicts the percentage of variation across studies due to heterogeneity rather than chance [I2 was used to assess heterogeneity, and I2 values of 0, 25, 50, and 75% represent no, low, moderate, and high heterogeneity respectively [The pooled measure was calculated as the inverse variance-weighted mean of the logarithm of RR with 95% CI. A random effects model was used to combine study-specific RR (95% CI), which considers both within-study and between-study variation . Statistn chance . The I2 ectively .P<0.05.Meta-regression with restricted maximum likelihood estimation  and subgA flow diagram of the study selection process was showed in Seven studies were included in our meta-analysis. The characteristics of the included studies on dietary zinc intake and the risk of pancreatic cancer are presented in I2=58.2%, P=0.026). In order to explore the significant between-study heterogeneity founded in the overall analysis, univariate meta-regression with the covariates of publication year, location (where the study was conducted), and study design (case–control or prospective) was performed. No significant findings were found in the above-mentioned analysis.The multivariate-adjusted RR of each study of the highest versus the lowest dietary zinc intake is available in P=0.997) also showed that there is no publication bias of the meta-analysis about dietary zinc intake and pancreatic cancer.Whether the result of the research has publication bias or not was showed in Sensitivity analysis  showed tIn this meta-analysis, highest category of dietary zinc intake was reversely associated with the risk of pancreatic cancer. We classified studies into several subgroups for analysis, of which the results were shown in I2=58.2%, P=0.026).Our study indicated that highest category of dietary zinc intake had significant statistical association on reducing the risk of pancreatic cancer. Those case–control studies and cohort ones were all of high quality. The results in case–control studies and American populations were consistent with the whole result. The total RR of pancreatic cancer for the highest versus the lowest categories of dietary zinc intake was 0.798 (0.621–0.984), with its significant heterogeneity among studies (Diet may be involved in the etiology of pancreatic cancer and dietary variations between countries may explain the differences in incidence of pancreatic cancer. For antioxidants, including dietary zinc intake, there are several plausible biological mechanisms by which they might prevent pancreatic cancer, including inactivating free radicals and reducing oxidative DNA damage, by stimulating immune function and through genetic effects ,22.This study reported the relationship between dietary zinc intake and pancreatic cancer risk with a comprehensive meta-analysis involving large number of cases and participants at the first time. The publication bias evaluated by Egger’s test and funnel plot had no significant association for the whole result or subgroup analyses, allowing a much greater possibility of reaching reasonable conclusions between dietary zinc intake and pancreatic cancer risk.However, evidence of significant between-study heterogeneity was found in the whole result and some subgroups analyses. As we all know, between-study heterogeneity is common in a meta-analysis, and exploring the heterogeneity is necessary in the report . TherefoHowever, our meta-analysis still had several restrictions. First, as our work embraced seven individual studies that were varied in quality, adjustments, and sample size that can cause the innate influence of our concluded result, but no publication bias was found. Second, due to the unconformity of categories of dietary zinc intake of each study, we did not do the dose–response analysis. Therefore, further studies with detailed category of dietary zinc intake are wanted to assess the dose–response analysis. Third, most studies followed a case–control design that lead to inherent recall and selection bias to retrospective studies. Although different kinds of studies were included, we did subgroup analysis to exclude the interruption. As only 2 prospective cohort studies involving 211 cases were included, more articles with cohort design are wanted.In summary, the present study suggested that highest category of dietary zinc intake had significant association on reducing the risk of pancreatic cancer, especially among American populations. During some limitation existed in our study, further studies with large cases and participants are wanted to confirm this result.


Pheterogeneity = 0.010), homozygote comparison , and the recessive genetic model  ; and in African descendants for homozygote comparison  and the recessive genetic model . In conclusion, the RNASEL Arg462Gln polymorphism may contribute to the risk of developing prostate cancer in African descendants and Hispanic Caucasians. Further larger and well-designed studies are warranted to evaluate this association in detail.The RNASEL gene  encodes a ribonuclease that plays a significant role in the apoptotic and antiviral activities of interferons. Various studies have used polymorphisms in the RNASEL gene to evaluate prostate cancer risk but studies that show an association between RNASEL Arg462Gln  polymorphism and prostate cancer risk are somewhat inconclusive. To assess the impact of RNASEL Arg462Gln polymorphism on prostate cancer risk, we conducted a meta-analysis of all available studies including 11,522 patients and 10,976 control subjects. The overall results indicated no positive association between the variant and prostate cancer risk. However, in a subgroup analysis by ethnicity, obvious associations were observed in Hispanic Caucasians for allelic contrast (OR = 1.18, 95% CI = 1.00 - 1.39,  Prostate cancer (PCa) is one of the most common types of neoplasm in the Western world. In United States, prostate cancer is the most prevalent cancer, with 217,730 new cases predicted to occur in 2010 , 5. PrioRibonuclease L (RNASEL), which is considered to be a tumor-suppressor gene, plays a significant role in the pathogenesis of prostate cancer through inflammation and infection. RNASEL is on chromosome 1q24-25 and encodes Ribonuclease L, a significant enzyme of the interferon-induced antiviral 2-5A system . MutatioExtensive epidemiological studies had been conducted to explore the association between RNASEL polymorphism and prostate cancer risk. A G-to-A transversion at nucleotide position 1385 (rs486907), which results in a glutamine instead of arginine at amino acid position 462 (R462Q), is one of the most widely investigated polymorphisms in RNASEL. Nevertheless, the association between the RNASEL R462Q polymorphism and prostate cancer risk is controversial because of conflicting case–control studies. Therefore, in this meta-analysis from all eligible studies published to date –32, we uA total of 22 articles (including 26 case–control studies) met all the inclusion criteria and were included Figure . The genPheterogeneity = 0.004, P = 0.758, I2 = 47.9), homozygote comparison , heterozygote comparison , the dominant genetic model , and the recessive genetic model. However, in the subgroup analysis by ethnicity, obvious associations between the RNASEL Arg462Gln polymorphism and prostate cancer risk were observed in African descendants for homozygote comparison , and the recessive genetic model ; and for Hispanic Caucasians for the recessive genetic model , homozygote comparison , and allelic contrast . No association was observed in Asian descendants for allelic contrast , homozygote comparison , and the recessive genetic model ; non-Hispanic Caucasians for allelic contrast  homozygote comparison , and the recessive genetic model ; and mixed descendants for allelic contrast , homozygote comparison , and the recessive genetic model   between the RNASEL polymorphism and hospital-based controls  and homozygote comparison , but not in sporadic prostate cancer for allelic contrast  and homozygote comparison .When all the eligible studies were pooled into the analysis Table , no posiP = 0.098; AA vs. GG, t = 1.77, P = 0.089; GA vs. GG, t = 1.75, P = 0.094; AA + GA vs. AA, t = 1.23, P = 0.231; AA vs. GA + GG, t = 1.77, P = 0.090)The Egger’s test and Begg’s funnel plot were carried out to assess the publication bias of the literature. No obvious evidence of publication bias was found  and Hispanic Caucasians , but not in Asian descendants and Non-Hispanic Caucasians. Furthermore, in a stratified analysis by source of control, the RNASEL Arg462Gln variant was found to increase prostate cancer risk in hospital-based studies (under the recessive genetic model). Nevertheless, several caveats limit generalization of these results. First, detailed information, such as age, prognostic parameters, environmental factors, and life-style, were not considered. Second, because different types of prostate cancer influence susceptibility, we tried to assess the effect of this polymorphism to different types of prostate cancer but not all data was compatible. Third, positive findings may be published faster than that those with “negative” results, which may result in a time-lag bias . In addiOther limitations of the meta-analysis need to be addressed. First, while it is possible that the Arg462Gln polymorphism contributes to cancer, the combined effects of multiple environmental or genetic components predominate in the development of carcinoma, and may mask the effect of the polymorphism . Second,In conclusion, this meta-analysis showed evidence that the RNASEL Arg462Gln polymorphism may contribute to the risk for developing prostate cancer in African descendants and Hispanic Caucasians, but not for other descendants. Further well-designed and prospective studies, particularly focused on gene-environment interactions, are warranted. These future studies should lead to a comprehensive understanding of the association between the RNASEL Arg462Gln polymorphism and prostate cancer risk.PubMed database searches were conducted using the following keywords: “ribonuclease L” or “RNASEL”, “prostate cancer”, and “polymorphism” . References of the relevant articles and retrieved paper were also screened by hand search. Eligible studies had to meet all the following criteria: (a) used an unrelated case–control design; (b) contained information about available genotype frequency; (c) published in English; and (d) included the full-text article.Data were collected on the genotype of rs486907 G/A (R462Q) according to prostate cancer. For each publication, the data extraction and methodological quality assessment were conducted by two of the investigators independently to ensure accuracy of the data. Disagreement was resolved by discussion between the two investigators. If they could not reach a consensus, the problem was discussed by all investigators to reach a consensus. The following parameters from each study were recorded: first author’s name, publication date, ethnicity, sources of cancer cases and controls, sample size in cases and controls, and the number of cases and controls with wild-type and variant allele, respectively.Q test. P value lower than 0.001 for the Q-test indicates lack of heterogeneity among studies, hence the pooled OR was utilized by the random effects model  [P values.Crude ORs with 95% CIs were utilized to evaluate the strength of association between the RNASEL polymorphism and prostate cancer based on genotype frequencies in cases and controls. Five genetic contrasts were used to assess the association: allelic contrast , homozygote comparison (AA vs. GG), heterozygote comparison (GA vs. GG), the dominant genetic model (AA + GA vs. GG), and the recessive genetic model (AA vs. GA + GG). Subgroup analysis was stratified by ethnicity, source of control , and smoking exposure. We utilized the random effects model and fixed effects model to calculate the pooled OR. Heterogeneity assumption was evaluated by a chi-square-based d method  or by thd method ). HWE waectively . We deteon bias) . All sta


P = 0.05, fixed-effect), and it could also predict poor OS  and RFS  in cancer patients. In conclusion, this meta-analysis indicated that MEG3 might serve as a potential novel biomarker for indicating the clinical outcomes in human cancers.Numerous studies have demonstrated that the expression level of maternally expressed gene 3 (MEG3) was lost in various cancers. Low expression of MEG3 is associated with an increased risk of metastasis and a poor prognosis in cancer patients. This meta-analysis investigated the association between MEG3 levels and distant metastasis (DM), lymph node metastasis (LNM), overall survival (OS), and recurrence-free survival (RFS) of cancer patients. A total of 536 participants from 9 articles were finally enrolled. The results showed a significant negative association between MEG3 levels and DM (OR = 2.16, 95% CI = 0.99–4.71,  Cancer has become one of the most important diseases threatening human health and life –3. ExploLong non-coding RNAs (lncRNAs) are a class of non-coding transcripts longer than 200 nucleotides , 10. By Maternally expressed gene 3 (MEG3) is the first lncRNA to be found to have tumor suppressor function, which is expressed in many human normal tissues . The losAs shown in Figure Table P = 0.23, fixed-effect) . Compared with high MEG3 expression group, low MEG3 expression group had a statistic significant elevated DM rate    Figure . Three s) Figure . Unfortu) Figure .I2 = 11%, P = 0.34), the fixed-effects model was chosen to estimate the pooled HRs with corresponding 95% CIs. Compared with high MEG3 expression group, low MEG3 expression group had a statistic significant reduced OS  , fixed-effects model was used. Analysis showed a pooled HR = 0.43 . Compared with high MEG3 expression group, low MEG3 had a statistic significant reduced RFS.We analysis pooled HRs of two group. Six studies with 384 patients were included in this meta-analysis of OS Figure . Because) Figure . Low MEGP > 0.05. Meanwhile, we used Stata11.0 software to evaluate sensitivity analysis to assess whether the individual studies affected the overall results. The results suggested that individual study had little influence on our final results . Furthermore, we found that there was significantly relationship between MEG3 and RFS . From these results, lncRNA MEG3 could be as biomarker for the prognosis of cancers. However, further large-scale studies should be conducted.MEG3 is an imprinted gene belonging to the DLK1-MEG3 locus located on chromosome 14q32.3 which is the first lncRNA to be found with tumor suppressor function . Previoun et al.  reportedP = 0.05). Unfortunately, there was no correlation in LNM ; insufficient sample size was the possible reason for the different results in this meta-analysis. Therefore, further studies should be done with larger sample sizes.It has been reported that MEG3 can play important role in inhibiting cancer through a variety of ways, which related to DNA methylation, P53 pathway, Rb pathway and so on . Sun L eIt should be stressed the limitations in our analysis. Most studies reported positive results, but those with negative results were generally less likely to be published. In addition, studies included in the meta-analysis most came from People's Republic of China, which might affect the results. Finally, there were insufficient data to fully confirm the association between MEG3 and clinicopathological characteristics, which needs more studies. Therefore, the results of this meta-analysis should be confirmed in future studies.In conclusion, despite the limitations described above, our meta-analysis reveals that the depressed expression of lncRNA MEG3 is significantly associated with DM, OS and RFS in patients with diverse cancers and could be a potential prognostic marker for cancers. However, large-scale and comprehensive researches were needed to illuminate our results. Well-designed studies related to specific cancer types and large sample sizes are needed to confirm the prognostic value of decreased lncRNA MEG3 in various cancers.We searched the databases PubMed, Cochrane library, Chinese National Knowledge Infrastructure, and Chinese Wan Fang database for studies published up to November 2016 to obtain relevant articles for the meta-analysis. The search strategy used both medical subject heading terms and free-text words to increase the sensitivity of the search. The keywords for the search were as follows: “MEG3 and cancer”, “long non-coding RNA MEG3”, “lncRNA MEG3”, “MEG3”. There was no language restriction. Meanwhile, reference lists of relevant articles were also reviewed to identify potential eligible papers.In this meta-analysis, eligible studies had to meet the following standards: 1)cohort design, 2) articles investigating the relation of MEG3 and cancer patients; 3) the expression levels of MEG3 in primary tumor tissues were measured, 4) sufficient original data for calculating odds ratios (ORs), hazard ratios (HRs) and their 95% confidence interval (95% CI). If the articles only provided survival curves without offering HR and 95% CI directly, appropriate data were extracted from the survival curves using Engauge Digitizer 4.1 software and the logHR and selogHR calculated according to Tierney et al. . If therTwo investigators extracted and reviewed relevant data from the eligible studies independently, including first author, year of publication, country, site of cancer, method, case number, cut-off value. If there were disagreements, a consensus was reached by the third investigator.Q-test and P-values. If heterogeneity was present (I2 ≥ 50% or P ≤ 0.05), random-effect model was used to calculate pooled HRs or ORs. If not, the fixed-effect model was more appropriate [P < 0.05 was considered statistically significant.Odds ratios (ORs) and 95% CI were used to evaluate the relationship between MEG3 and in these inclusive articles. The features included gender, distant metastasis (DM), and lymph node metastasis (LNM). Meanwhile, HRs and 95% CIs were used to assess the association between MEG3 and cancer prognosis . We used Revman5.3 Software  to perform the meta-analysis and evaluate heterogeneity between studies by Cochrane ropriate , 46. The


Coffee consumption has been associated with numerous cancers, but evidence on ovarian cancer risk is controversial. Therefore, we performed a meta-analysis on prospective cohort studies in order to review the evidence on coffee consumption and risk of ovarian cancer.Studies were identified through searching the PubMed and MEDLINE databases up to March 2017. Risk estimates were retrieved from the studies, and dose-response analysis was modelled by using restricted cubic splines. Additionally, a stratified analysis by menopausal status was performed.A total of 8 studies were eligible for the dose-response meta-analysis. Studies included in the analysis comprised 787,076 participants and 3,541 ovarian cancer cases. The results showed that coffee intake was not associated with ovarian cancer risk . Stratified and subgroup analysis showed consisted results.This comprehensive meta-analysis did not find evidence of an association between the consumption of coffee and risk of ovarian cancer. Coffee is among the most consumed beverages worldwide: in Europe, the geographic area showing the highest coffee consumer in the world, people are reported to drink 725 million cups of coffee every day . In lighCurrent evidence from epidemiological and experimental studies suggests that coffee consumption may exert beneficial effects towards non-communicable diseases . In partOvarian cancer has a major epidemiological and social burden for women worldwide: in 2012, women with a diagnosis of ovarian cancer occurred in the last 5 years were over 587,000 and 157,000 worldwide and in Europe, respectively , 19. In The evidence on the association between coffee drinking and the risk of ovarian cancer is quite limited and inconsistent. Results of a meta-analysis of prospective cohort studies revealed a non-significant positive association between coffee drinking and risk of ovarian cancer . HoweverWe performed a meta-analysis of prospective cohort studies following Meta-Analysis of Observational Studies in Epidemiology (MOOSE) protocols throughout design, execution, analysis and reporting stages .http://www.ncbi.nlm.nih.gov/pubmed/) and EMBASE (http://www.embase.com/) databases, from the earliest available online indexing year to March 2017. The search was limited to studies published in English. Search terms included the following: (coffee OR caffeine OR beverages) AND (ovarian) AND (cancer OR carcinoma OR neoplasm)  . Titles Data were abstracted from all identified studies using a standardized extraction form. Information was extracted from each study and consisted of: 1) first author name; 2) year of publication; 3) study cohort name; 4) country; 5) number of participants; 6) gender of participants; 7) age range of the study population at baseline; 8) categories of consumption; 9) coffee type; 10) follow-up period; 11) endpoints and cases; 12) distributions of cases and person-years, HRs, and 95% CIs for all categories of exposure; 13) covariates used in adjustments. Extraction of data was performed independently by two authors. Discrepancies were discussed and resolved by consensus. The quality of included studies was assessed according to the Newcastle-Ottawa Quality Assessment Scale .versus lowest and dose-response meta-analyses were performed to determine the relationship between coffee intake and ovarian cancer risk. The most fully adjusted RRs/HRs with 95% CI for all categories of exposure were extracted. Random-effects models were used to calculate pooled effects, wherein HRs were deemed as equivalent to relative risks (RRs) [I2 statistic, which estimates the fraction of variance that is due to heterogeneity and by Q test. The level of significance for the Q test was defined as P < 0.10. The values of I2 statistic ≤25%, 25–50%, 50–75%, and >75% indicated no, small, moderate, and significant heterogeneity, respectively. The stability of the results and potential sources of heterogeneity were examined in a sensitivity analysis excluding one study at a time and in subgroup analysis according to menopausal status and geographical area. Included studies did not provide datasets of stratified analysis to test for potential confounders/effect modifiers, such as smoking or BMI. Therefore, subgroup analyses were only performed according to adjustment for smoking status, BMI, education level and alcohol consumption. Publication bias was evaluated by a visual investigation of funnel plots for potential asymmetry.Highest ks (RRs) . HeterogP-value by testing that the 2 regression coefficients were simultaneously equal to zero. We then calculated a P-value for non-linearity by testing that the coefficient of the second spline was equal to zero. All analyses were performed with R software version 3.0.3 and we mainly used dosresmeta package .To examine linear and non-linear relationship between coffee intake and risk of ovarian cancer, random-effects dose-response meta-analysis was used. We extracted data on the amount of coffee intake, distributions of cases and person-years (when available), and RRs/HRs with 95% CIs for ≥3 exposure categories. In each study across categories of consumption, the mean or median intake of coffee was assigned to the corresponding RR/HR with the 95% CI. When originally in article the range of coffee intake was reported , the midpoint of the range was selected . When the highest category was open ended, we assumed the width of the category to be the same as the adjacent category. When the lowest category was open ended, we set the lower boundary to zero. Random-effects dose-response meta-analysis was performed in two-stages. In the first stage the method reported by Greenland and Orsini  was used to calculate study-specific slopes  on the basis of results across categories of coffee intake taking into account the correlation within each set of retrieved HRs , 25. ModWe identified a total of 1,340 studies, of which 1,223 were excluded after review of title, and 104 on the basis of abstract , which in turn could be related to higher risk of cancers due to a synergistic effect of many functional components rather than an individual food or beverage . MoreoveA major strength of our meta-analysis was the inclusion of cohort studies carried out with a prospective design, which implies detailed exposure and covariate assessment before the diagnosis of the outcome of interest . Moreover, we performed a dose-response meta-analysis, which scientific value is higher than meta-analyses with exposure classified as high versus low, as it aims to investigate a possible dose-response relationship. However, the results of this study should be considered in light of some limitation. First, no stratified analysis was performed due to lack of available datasets, which prevented us from an in-depth analysis of potential confounders and effect modifiers, such as smoking status or body weight; despite our adjustment for smoking status, BMI index, education level and alcohol exposure, there is still a chance of unmeasured or residual confounding . Another limitation lies in the exposure assessment phase: the categories of low and high consumption varied across different studies, therefore overall estimates of high consumption might not be perfectly comparable; moreover, we don’t have data on coffee brewing methods, preparation, cup size, and duration of consumption. Such differences, if properly addressed, could yield to significant results in specific exposure subgroups  . SimilarIn conclusion, the present study provided a robust assessment of the relationship between coffee consumption and ovarian cancer, which, coherently with previously published literature, appears null. Additional prospective cohort studies specifying also subgroup analyses by key variables  are needed to further improve current knowledge on such topic.


Dairy product consumption may affect the risk of hip fracture, but previous studies have reported inconsistent findings. The primary aim of our meta-analysis was to examine and quantify the potential association of dairy product consumption with risk of hip fracture.We searched the databases of PubMed and EMBASE for relevant articles from their inception through April 17, 2017. The final analysis included 10 cohort studies and 8 case-control studies. Random-effects models were used to estimate the pooled risk. Subgroup and dose-response analyses were conducted to explore the relationships between the consumption of milk and the risk of hip fracture.After pooling the data from the included studies, the summary relative risk (RR) for hip fracture for highest versus lowest consumption were 0.91 (95% CI: 0.74–1.12), 0.75 (95% CI: 0.66–0.86), 0.68 (95% CI: 0.61–0. 77), 1.02 (95% CI: 0.93–1.12) for milk, yogurt, cheese, and total dairy products in cohort studies, respectively. Higher milk consumption  was associated with lower risk of hip fracture for highest versus lowest consumption in case-control studies. After quantifying the specific dose of milk, the summary RR/OR for an increased milk consumption of 200 g/day was 1.00 (95% CI: 0.94–1.07), and 0.89 (95%CI: 0.64–1.24) with significant heterogeneity for cohort and case-control studies, respectively; There was a nonlinear association between milk consumption and hip fracture risk in cohort, and case-control studies.Our findings indicate that consumption of yogurt and cheese was associated with lower risk of hip fracture in cohort studies. However, the consumption of total dairy products and cream was not significantly associated with the risk of hip fracture. There was insufficient evidence to deduce the association between milk consumption and risk of hip fracture. A lower threshold of 200 g/day milk intake may have beneficial effects, whereas the effects of a higher threshold of milk intake are unclear.The online version of this article (10.1186/s12889-018-5041-5) contains supplementary material, which is available to authorized users. Hip fracture is the most serious type of osteoporotic fracture. Hip fractures can lead to other comorbidities, increased mortality risk, and enormous social and economic costs . AccordiDairy products have been hypothesized to help prevent hip fracture because they are a significant source of calcium, proteins, and other bioactive nutrients beneficial for bone health . HoweverTwo previous meta-analyses have been published with results focused only on milk consumption , 15. HowThe primary aim of our meta-analysis was to examine and quantify the potential association of dairy product consumption with risk of hip fracture. Searches were performed on PubMed and EMBASE databases from their inception to April 17, 2017. The search terms used were as follows: “hip fracture”  and “dairy products” (or “milk” or “cheese” or “yogurt” or “cream”). No language restrictions were applied in the search strategy. An additional article  was idenTo identify eligible studies, two independent investigators (S.B.and J.H.) performed an initial screening of all titles and abstracts, and then assessed the full text of all relevant studies in detail. Articles were included in this meta-analysis if they met the following criteria: 1) cohort or case-control study design; (2) studies that evaluated and clearly defined exposure to total dairy products, milk, cheese, yogurt, or cream; (3) studies with the risk of hip fracture as the outcome of interest; and (4) studies reporting odds ratio (OR) or relative risk (RR) with 95% confidence intervals (CI) for the association between dairy product exposure and risk of hip fracture. Studies had to define hip fracture using the criteria based on the International Classification of Diseases, 10th revision (ICD-10) or medical records. If there were multiple publications from the same study, we selected the most recent study for the meta-analysis. Studies were excluded if they provided insufficient data, such as letters, reviews, comments, or animal studies. Two previous meta-analyses had been published with results focused only on milk consumption. The meta-analysis conducted by Bischoff-Ferrari et al.,  was excl cohort oTwo authors (Y.W. and K.Z.) independently extracted the following information from each included study: author name, research region, publication year, study design, study name, subjects (number of cases), sex, mean/median age of the study individuals, duration of follow-up for cohort studies, exposure and quantity of intake, dietary assessment method, the maximally adjusted risk estimates with 95% CI for the highest versus the lowest category of consumption, and adjustment for confounders in analyses.In the dose-response meta-analysis of the relationship between dairy products and hip fracture risk, the number of cases and participants or person-years, the mean or median dairy product consumption for each exposure category, and the RR/OR and its variance estimate for three or more quantitative exposure categories were compiled from the included studies. The median level of milk consumption (g/day) for each exposure category was presented with the relevant RR/OR and corresponding 95% CI. We used standard conversions from the Food Standards Agency to convert glasses/d to g/d (1 glass = 200 mL) for relevant studies , 18. AssTwo investigators (M.Y. and K.Z.) independently assessed the quality of the 17 included studies , Australia (n = 1), USA (n = 1), and Asia (n = 3).Table Nine cohort studies , 37, 42 2 = 75.0%, Pheterogeneity < 0.01)  Fig. , indicatP = 0.81) and Begg’s test (P = 0.30) indicated no publication bias with regard to milk intake and hip fracture risk. We used the trim-and-fill method to confirm robustness of the results. There were no significant changes to the results after using the trim-and-fill method when including four missing articles .The contour-enhanced funnel plot demonstrated asymmetry Fig. . However2 = 54%, Pheterogeneity = 0.04)  Fig. . Pooled For case-control studies, the contour-enhanced funnel plot showed asymmetry Fig. . There wThe quality of the 17 included studies  [nonlinerity < 0.01), with a rapid increase in risk when milk consumption increased from 0 to 600 g/d; there was no further increase in risk with milk consumption between 600 and 1200 g/d.The relationship between milk consumption and hip fracture risk was further quantified via dose-response analysis for milk consumption. The summary RR for an increased milk consumption of 200 g/day was 1.00 (95% CI: 0.94–1.07), with significant heterogeneity among cohort studies (I. n = 7) , 37, 42 2 = 68%, Pheterogeneity = 0.05, Fig. n = 3) [nonlinerity = 0.28), with a reduction in risk with milk consumption of 200–600 g/d. However, the confidence intervals were wide for all outcomes , with significant heterogeneity among case-control studies (I. n = 3) , 39, 41 mes Fig. .n = 3) and cheese (n = 3) reduced hip fracture risk, total dairy products (n = 2) and cream (n = 1) showed no association with hip fracture risk in cohort studies for the highest versus lowest category , yogurt (n = 1), and cheese (n = 3) in case-control studies showed no association with hip fracture risk  PubMed, and b) EMBASE. (DOC 37 kb)Additional file 2:Title: Selection procedure for inclusion and exclusion of the studies. (DOC 50 kb)


FSHB)  and follicle stimulating hormone receptor (FSHR)  genes might disturb normal spermatogenesis and affect male reproductive ability.Male infertility is a complex disorder caused by genetic, developmental, endocrine, or environmental factors as well as unknown etiology. Polymorphisms in the follicle stimulating hormone beta subunit  evaluation, no statistically-significant difference was found in the frequency of allelic variants or in genotype distribution between cases and controls. However, a significant association for the comparison of GAA  was seen between the oligozoospermia and controls in haplotype analysis of rs1394205/rs6165/rs6166. In the meta-analysis, rs6165G allele and rs6166 GG genotype were associated with increased risk of the male infertility.In the FSHR GAA haplotype would exert protective effects against male sterility, which indicated that the combination of three SNP genotypes of FSHR was predicted to have a much stronger impact than either one alone. Then in the meta-analysis, a significant association was seen between FSHR rs6165, rs6166 polymorphisms and male infertility. In terms of male infertility with multifactorial etiology, further studies with larger sample sizes and different ethnic backgrounds or other risk factors are warranted to clarify the potential role of FSHB and FSHR polymorphisms in the pathogenesis of male infertility.This study suggested that  Worldwide, approximately 15% of couples cannot have a child, and in half of the cases the reason is the result of male infertility . It has MTHFR) , dominant model (RR + WR vs. WW), recessive model (RR vs. WW + WR). In addition, stratified analyses were performed by HWE (>0.05 and <0.05), race  and case counts (>200 and <200). HWE < 0.05 indicates unbalanced distribution in the frequency of the population. So studies with SNPs with HWE < 0.05 were removed from meta-analysis. Heterogeneity across the studies was evaluated by t-test test based on Q test and was considered significant if P-value for heterogeneity (Ph) was <0.05. A fixed-effect model with no heterogeneity (Ph > 0.05 or I2 < 50%) using the Mantel–Haenszel method and a random effects model with a high heterogeneity (Ph < 0.05 or I2 > 50%) using the DerSimonian and Laird method were used to pool the results. Moreover, a sensitivity analysis, by which a single study in the meta-analysis was deleted each time to determine the influence of the individual data set to the overall pooled OR, was performed to assess the stability of the results. To test the publication bias, Begg’s Funnel plots and Egger’s linear regression test were applied. HWE in the controls of each study was calculated using a web-based program (http://ihg.gsf.de/cgi-bin/hw/hwa1.pl).This study and other related case-control studies were combined for the meta-analysis, which was performed using STATA 12.0 . Odds ratio (OR) and 95% confidence interval (CI) were calculated to estimate the associations of P: 0.538 for FSHB rs10835638, P: 0.975 for FSHR rs1394205, P: 0.884 for rs6165, P: 0.391 for rs6166, respectively). The serum FSH and LH concentrations in fertile men were 4.72 ± 2.51 IU/L and 3.37 ± 1.46 IU/L, whereas the FSH and LH values in infertile patients were 15.73 ± 16.15 IU/L and 6.46 ± 4.98 IU/L, respectively. The FSH and LH levels in the infertile patients were significantly higher than that in the fertile men (P < 0.05). And fertile men had higher sperm concentration and sperm motility compared with the infertile men. No statistical difference was observed in the other indicators.A total of 255 infertile men and 340 infertile men were collected in this case-control study. The clinical charactereristics of the participants were presented in Table FSHR SNPs at nucleotide −211, −29, codon 307 and codon 680 was separately analyzed, no statistically significant difference was found in the frequency of allelic variants or in genotype distribution between cases and controls, as showed in Table Logistic regression analysis revealed that, when the P: 0.022, OR: 0.63, 95%CI: 0.43–0.94) was seen between the oligozoospermia and controls in haplotype analysis of rs1394205/rs6165/rs6166  between SNPs 307/680 , weak LD between SNPs −29/307 , and SNPs −29/680  locuses. A significant association for the comparison of GAA . In addition, in the subgroup analysis with HWE > 0.05, significant differences was observed for the genotype GG , AG  and GA + GG . By the race, one paper which reported in the Brazilian population, revealed a significant association for the comparison of AG vs. AA , GA + GG vs. AA . And a slight significant association for the AG vs. AA  was found in the subgroup analysis of case count > 200. However, no significant association was revealed in the remaining subgroup analyses, as summarized in Table For rs6165 polymorphism, the overall analyses showed that AG genotype was associated with increased risk of male infertility , GG vs. AA + AG  in subgroup analysis with HWE > 0.05, and the comparison of GG vs. AA  in subgroup analysis with case count >200, as presented in Table For rs6166 polymorphism, no significant association was observed under the all genetic models in overall analyses; however, significant differences was showed for the comparison of GG vs. AA  in overall analysis, homozygote model (GG vs. AA: Ph: 0.038), dominant model (GA + GG vs. AA: Ph: 0.044) in subgroup analysis of case count > 200. The heterogeneity was decreased respectively to 0.120, 0.266 and 0.201 when omitting the paper reported by Gharesi-Fard et al. [Sensitivity analyses on the d et al. , was theFSHR rs6165, rs6166 and rs1394205 funnel plot shape did not show any evidence of obvious asymmetry in all comparison models. And the Egger’s test was used to provide statistical evidence for funnel plot symmetry, which showed no obvious evidence of publication bias.Begg’s Funnel plots and Egger’ test were performed to assess publication bias. For the FSHB, FSHR gene polymorphisms and male infertility. Compared with the fertile controls, the infertile patients had higher FSH and LH levels, and lower sperm concentration and sperm motility. Then the most common FSHR allelic variants in the core promoter and exon 10, and the FSHB variants in the core promoter were genotyped with respect to male infertility status using the Mass ARRAY platform. However, the distributions of FSHB and FSHR allele, genotype frequencies among azoospermic, severe oligozoospermic, or oligozoospermic men and fertile men based on age-adjusted estimates were similar. However a more precise analysis should be conducted if all individual raw data were available, to allow for the adjustment according to hormone, sperm parameters and other lifestyle factors. Consistent with previous reportes [P: 0.022, OR: 0.63, 95%CI: 0.43–0.94) among the oligozoospermic men in haplotype analysis, which indicated that the GAA haplotype would exert protective effects against male sterility.In the present study, we included 340 fertile males and 255 infertile males, consisting of 166 with azoospermia or severe oligozoospermia, and 89 with oligospermia, and we investigated the association between reportes , 34, we FSHR polymorphisms and male infertility parameters have produced contradictory results. Variants of FSHR have been shown to affect the serum FSH, inhibin B, anti-Mullerian hormone (AMH) and total testes volume [FSHR isoforms and male reproductive parameters. It is regrettable that not all clinical fertility parameters in this study were collected, as we could not find any association between the clinical fertility parameters and the FSHB and FSHR genotypes.Previous studies examining potential associations between s volume , 25, 33.FSHB -211G > T T-allele showed significant dosage effects on FSH, LH and bilateral testicular volume. Moreover, FSHR 2029A > G significantly modulated the more dominant effect of FSHB -211G > T on serum FSH and testicular volume among the T-allele carriers [FSH secreted by the anterior pituitary, together with other endocrine factors, plays a central role in establishing and maintaining human fertility. In males, circulating FSH stimulates gametogenesis and steroidogenesis in the gonads by binding to its receptor (FSHR). Tuttelmann et al. reported that the carriers , suggestFSHR polymorphisms and male infertility risk.We searched available databases, such as GWAS Central, National Human Genome Research Institute GWAS Catalog and PUBMED, but failed to find a relevant genome-wide association study (GWAS) on all SNPs. To date, no clear consensus appears to have been reached in the literature on the relationship between FSHR rs1394205 (G-29A), rs6165 (Thr307Ala), or rs6166 (Asn680Ser) polymorphisms and the risk of male infertility [FSHR rs1394205 polymorphism was not associated with male infertility, which was similar to the findings of other studies.To resolve the conflicting results, we carried out a meta-analysis to obtain a more precise estimation of the associations. Contrary to previous meta-analyses that found no significant association between s6165 Thr7Ala, or ertility , this stFSHR rs1394205 polymorphisms and male infertility. Sensitivity analysis revealed that the study reported by Li et al. [FSHR rs1394205A allele and male infertility, resulting in the significant heterogeneity.In addition, identifying the source of heterogeneity is one of the most important goals of the meta-analysis. In each case, the heterogeneity could be a result of differences in ethnicity, sources of controls, methods used and so on. In this study, we found that significant heterogeneity existed in overall analysis and sub-group analysis for the comparison of i et al. , containi et al. , comprisi et al. . HoweverIt should be pointed out that there are some limitations in this meta-analysis. Firstly, inadequate sample size and inappropriate control subjects resulted in limited power for exploring the real association, and subgroup analyses by HWE, ethnicity, and case count involved relatively small groups, which were more likely to reveal greater beneficial effects than a large-scale trial. Secondly, owing to lack of the original data, we could not further evaluate the potential interactions between genes, which might affect male infertility. Thirdly, much detailed information, including body mass index, age, work, smoking or alcohol habits, environmental exposure and other lifestyle factors, is not available, so that our results were based on unadjusted estimates. A more precise analysis should be conducted through adjustment by other co-variants. In fact, some other genes as well as environmental exposure could also play an important role in spermatogenesis.FSHB and FSHR genes might disturb normal spermatogenesis and affect male reproductive ability. Because of this, FSH treatment has always been tempting and is actively prescribed by many doctors even though the efficacy of FSH therapy remains a contentious issue. One study showed that patients with at least one FSHR c.2039A > G G allele had a significant increase in total sperm count after 3 months of treatment with recombinant FSH . Another study considered the FSHB-211G > T genotype and showed that TT homozygote, representing 25% of men with oligozoospermia and low FSH levels, could significantly benefit from FSH treatment [FSHB and FSHR genotypes [Owing to the critical role of FSH in spermatogenesis, polymorphisms in the reatment . Howeverenotypes .FSHR GAA haplotype would exert protective effects against male sterility, which indicated that the combination of three SNP genotypes of FSHR was predicted to have a much stronger impact than either one alone. Then in the meta-analysis, a significant association was seen between FSHR rs6165, rs6166 polymorphisms and male infertility. In terms of male infertility with multifactorial etiology, further studies with larger sample sizes and different ethnic backgrounds or other risk factors are warranted to clarify the potential role of FSHB and FSHR polymorphisms in the pathogenesis of male infertility.This study suggested that