Background: We performed a systematic review and meta-analysis to evaluate the risk of pneumonitis and pneumonia associated with immune checkpoint inhibitors (ICIs) for solid tumors.Methods: The following keywords were used in searching the Embase and PubMed database: pneumonitis, pneumonia, and immune checkpoint inhibitors. The data was analyzed by using the R software and Metafor package.Results: Among 3,436 studies, 23 randomized clinical trials (RCTs) met our selection criteria which included data from 12,876 patients. Compared with chemotherapy, PD-1 inhibitors showed significant increase in grade 1-5 and grade 3-5 pneumonitis , but not in pneumonia. PD-L1 inhibitors showed significant increase in grade 1-5 pneumonitis and pneumonia . There was no significant difference in any grade pneumonitis and pneumonia in cytotoxic T lymphocyte-associated protein 4 (CTLA4) inhibitors subgroup. Programmed cell death protein 1 (PD-1) inhibitor (nivolumab and pembrolizumab) both showed significant increase in grade 1-5 pneumonitis, and pembrolizumab specially tended to increase grade 3-5 pneumonitis. . Compared with PD-1 inhibitor (nivolumab) or CTLA-4 inhibitor (ipilimumab) monotherapy, PD-1 inhibitor, and CTLA-4 inhibitor (nivolumab plus ipilimumab) combination therapies showed significant increase in grade 1-5 and grade 3-5 pneumonitis .Conclusions: PD-1/PD-L1 inhibitors treatment could increase the risk of all-grade pneumonitis. CTLA4 inhibitor ipilimumab treatment alone could not increase the risk of pneumonitis but could augment the risk of pneumonitis in PD-1/PD-L1 inhibitor treated patients. There was no significant increase in the risk of pneumonia after either PD-1/PDL-1inhibitor or CTLA4 inhibitor treatment alone or in combination. AwareneUsing the search terminology, we initially identified 3,436 studies from our database search. Among those 3,436 studies, 23 RCTs met our strict inclusion criteria . All theCochrane risk of bias tool was used to measure the quality of the included studies and the results are shown in pneumonitis ,but not in pneumonia. Compared with control, PD-L1 inhibitors showed significant increase in grade 1-5 pneumonitis and pneumonia . There was no significant difference in any grade pneumonitis and pneumonia in CTLA4 inhibitors subgroup.As shown in p < 0.001; RR 3.48, 95%CI:1.10-11.02, p < 0.001) (Compared with PD-1 inhibitor (nivolumab) or CTLA-4 inhibitor (ipilimumab) monotherapy, PD-1 inhibitor and CTLA-4 inhibitor (nivolumab plus ipilimumab) combination therapies showed significant increase in grade 1-5 and grade 3-5 pneumonitis  , but not< 0.001) .Compared with chemotherapy, there was no significant difference in the risk of any grade pneumonitis and pneumonia in ICIs plus chemotherapy combination therapies. When compared with CTLA-4 inhibitors, the risk of any grade pneumonitis induced by PD-1inhibitors seems higher, which however, was not statistically significant .pneumonitis , but only pembrolizumab showed significant increase in grade 3-5 pneumonitis. , while nivolumab did not show significant increase . There was no significant difference in grade 1-5 and grade 3-5 pneumonia in nivolumab or pembrolizumab subgroup. PDL-1 inhibitor Atezolizumab showed significant increase in grade 1-5 pneumonitis and pneumonia .Compared with control, there was no significant difference in grade 1-5 or grade 3-5 pneumonitis in CTLA-4 inhibitor ipilimumab subgroup.As shown in pneumonitis and pneumonia) between any ICI treatment group and control treatment group  in NSCLS subgroup. There was no significant increase in the risk of pneumonia in PD-1 inhibitor treated patients. Similarly, in other tumor types including MM, HNSCC, prostate cancer and mesothelioma, PD-1 inhibitors showed significant increase in grade 1-5 and grade 3-5 pneumonitis  but no increase in the risk of pneumonia. In NSCLS subgroup, compared with control therapeutics, PD-L1 inhibitors showed significant increase in grade 1-5 pneumonitis and pneumonia . There was no significant difference in the risk of any grade pneumonitis and pneumonia with CTLA4 inhibitors in other tumor types.As shown in pneumonitis and pneumonia incidence in all the remaining subgroups .There was very tiny overall heterogeneity of grade 1-5/3-5 P-values were > 0.05 after both tests. Therefore, there was no significant publication bias in this meta-analysis.Egger's test and Begg's test, conducted by STATA 12.0 software, were utilized to evaluate the publication bias between different RCTs. As presented in When combating some solid tumors, ICI therapy alters the balance of immune cells in the body, which in turn cause damage to certain organ system, this phenomenon called immune-related adverse events (irAE) such as pneumonitis, colitis, and endocrine disorders , 26. In Our meta-analysis demonstrated that PD-1/ PD-L1 inhibitors including nivolumab, pembrolizumab, and Atezolizumab could increase the risk of all-grade pneumonitis, which was not seen in CTLA4 inhibitor (ipilimumab). Only pembrolizumab showed significant increase in grade 3-5 pneumonitis compared to chemotherapy. This result is consistent with other reports in that pneumonitis appears to be more common with PD-1/PD-L1 inhibitors , 31. SevOur meta-analysis showed that when combining ipilimumab (CTLA-4 inhibitors) with nivolumab (PD-1 inhibitors), the risk of all-grade pneumonitis (3.47 times), and severe pneumonitis (3.48 times) were higher than nivolumab or ipilimumab alone. Thus, the combined inhibitors of CTLA-4 and PD-1 could generate higher incidences of pneumonitis than either blockade –23. It wIn this study, we found that PD-L1 inhibitor increased the risk of pneumonia compared to chemotherapy/placebo. However, we did not observed any significant difference in the risk of pneumonia for PD-1 and CTLA4 inhibitors monotherapy (vs. chemotherapy), and combination therapy (vs. monotherapy). Interestingly, patients with PD-L1 inhibitors treatment for NSCLC had a higher incidence of pneumonia than those patients with other cancer types. In one American analysis of data from 11,111 lung cancer patients and 49,975 patients with other solid tumors, pneumonia occurred more commonly in the lung cancer patients (26.4 vs. 10.3%) . TherefoIn reality, both pneumonitis and pneumonia were uncommon. Potentially serious lung toxicities occur in only 1–7 and 1–5% of ICI-treated patients, respectively. Most cases of pneumonitis and pneumonia were mild. With the increasing use of ICIs in many anticancer settings, the absolute burden and mortality of pneumonitis and pneumonia will undoubtedly rise. This meta-analysis study provided new information to physicians regarding the difference of pneumonitis and pneumonia in cancer patients treated with ICIs, especially those with PD-1 and CTLA4 inhibitors. Further study on the molecular mechanisms underlying these side effects of ICI therapy could help us to implement a better therapeutic strategy and to avoid some of these side effects .This type of meta-analysis itself based on published data had several unavoible limitations. First, this meta-analysis did not include individual patient data, the use of which would have provided more details about pneumonitis and pneumonia with ICIs. Secondly, the sensitivity analysis was not employed in this meta-analysis because of the paucity of the study number on pneumonia with nivolumab and ipilimumab vs. ipilimumab. Thirdly, the clinical studies included were not specifically designed to assess the immune related pneumonitis and pneumonia, and a generally acknowledge of the diagnostic criteria is still lacking. Cancer patients are generally prune to have lung infection due to overall health condition and the compromised immune system. Some lung infections or tumor progression could have been misdiagnosed clinically as pneumonitis. Therefore, the identification of autoimmune pneumonitis in these studies may not be completely accurate or homogeneous. Moreover, the final diagnosis of immune pneumonitis was not central reviewed, so it could depend on the experience of every center. Prospective centrally reviewed multicenter studies with more stringent diagnostic criteria could help us better understand the relative risk and the pathogenesis of these immune related pulmonary complications. On the other hand, we have made great efforts on the overall quality assessment to make our conclusion more steady and credible: (1) two independent reviewers searched all the relevant trails with well-defined inclusion criteria. They assessed studies appropriate for meta-analysis evaluated by using PICO chart and assessed the risk of bias for the included RCTs according to the Cochrane Handbook. (2) Two independent reviewers verified data in our meta-analysis which was performed by pair-wise comparisons. (3) The random-effects model and subgroup analysis were employed statistically in this meta-analysis. (4) The heterogeneity of nearly all the subgroups was low or moderate.In summary, our meta-analysis has demonstrated that PD-1/ PD-L1 inhibitors including nivolumab, pembrolizumab and atezolizumab could increase the risk of all-grade pneumonitis rather than CTLA4 inhibitor (ipilimumab). Only pembrolizumab showed the significant increase in grade 3-5 pneumonitis compared to chemotherapy. The risk of Grade 1-5/3-5 pneumonitis of combined ipilimumab with nivolumab was higher than nivolumab/ipilimumab alone. PD-L1 inhibitor may increase the risk of pneumonia compared to chemotherapy/placebo. Clinicians need to be aware of these ICI-associated respiratory disorders when employing ICI therapy for solid tumors so that patients can be appropriately managed.The systematic review with meta-analysis was conducted according to the guidelines of the Cochrane Handbook for Systematic Reviews of Interventions , and rephttps:// clinicaltrials.gov.  for studies reporting the risk of pneumonitis or pneumonia associated with ICIs monotherapy vs. chemotherapy or control, combination therapy (PD-1 inhibitor plus CTLA4 inhibitor) vs. ICI single therapy, and combination therapy (ICIs plus chemotherapy) vs. chemotherapy for the treatment of patients with solid tumors. The medical subject heading (MeSH) terms included in searching the relevant studies contained one term that means neoplasms , one term means ICIs , and one term related to randomized controlled trials (RCTs). We used “and” to connect the terms  Phase II/III RCTs with primary endpoints including overall survival (OS), progression-free survival (PFS), or objective response rate (ORR); (2) histologically confirmed solid cancer such as lung cancer, and others; (3) containing the information of ICIs , controlled therapies, pneumonitis, and pneumonia.The studies were excluded if they were: (1) letters, reviews, unfinished studies, duplicate reports, or conference reports; (2) studies conducted with animal models or cell lines; (3) studies due to insufficient data; (4) papers in other languages than English; (5) RCTs in phase I.Y.L.H. and Q.S.) searched all the relevant studies and read the titles, abstracts, and full texts of the identified studies. We accessed each study appropriateness for meta-analysis by using the PICO  chart ((X.C.Z.).Two independent reviewers (e) chart . The fole) chart . DisagreQ.S. and Y.L.H.) independently assessed the quality of the included RCTs according to the Cochrane risk of bias tool, which assesses the following seven domains: selection bias , performance bias, detection bias, attrition bias, reporting bias, and other bias. R3.4.3 (R Project). The metafor package software was used for our meta-analysis. The Risk Ratio (RR) was used to estimate overall and severe pneumonitis or pneumonia (grade 1-5/3-5/5). RR >1.0 indicates higher risk or higher incidence of overall and severe pneumonitis or pneumonia in patients treated with ICIs than those treated with chemotherapy or placebo. In addition, the Q test and I2 statistics were used to assess the heterogeneity among the RCTs. I2 values of <30, 30–59, 60–75, and >75% were classified as low, moderate, substantial, and considerable heterogeneity, respectively. . Two reviewers . Subgroup analysis and sensitivity analysis were used to assess the heterogeneity among the RCTs. Random-effects model (REM) was employed to validate the statistical results in our meta-analysis.PICO chart was used to assess study's appropriateness for meta-analysis. Cochrane Handbook for Systematic Reviews of Interventions was used to assess the risk of bias for the included studies. All disagreements in our meta-analysis were resolved by discussion with the third reviewer (QS and EZ had access to all the data included in the study and are responsible for the completeness of the data and the accuracy of our analysis. JW, YH and TL helped to design the study. QS, DW, and YH contributed to the statistical analysis and the revision of this manuscript. QS and ZG approved the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


CYP1A1 rs1048943, EGF rs444903, HOTAIR rs920778, MMP2 rs243865, and PLCE1 rs2274223, 10 variants were rated as moderate, and 18 variants were rated as weak. Additionally, 17 SNPs were verified noteworthy in six genomewide association studies (GWAS) using FPRP methods. Collectively, this review offered a comprehensively referenced information with cumulative evidence of associations between genetic polymorphisms and EC and ESCC risk.An increasing number of publications had reported the association between single‐nucleotide polymorphisms (SNPs) and esophageal cancer (EC) risk in the past decades. Results from these publications were controversial. We used PubMed, Medline, and Web of Science to identify meta‐analysis articles published before 30 July 2018, that summarize a comprehensive investigation for cumulative evidence of genetic polymorphisms of EC and its subtype risk. Two methods, Venice criteria and false‐positive report probability (FPRP) tests, were used to assess cumulative evidence of significant associations. At last, 107 meta‐analyses were considered to be in conformity with the inclusion criteria, yielding 51 variants associated with EC or esophageal squamous cell carcinoma (ESCC). Thirty‐eight variants were considered to be nominally significant associated with risk of EC or ESCC, whereas the rest showed non‐association. In additional, five variants on five genes were rated as strong cumulative epidemiological evidence for a nominally significant association with EC and ESCC risk, including  Ioannidis et al indicated that mechanisms for summarizing and assessing genetic epidemiological evidence require periodic updates of all appropriate association studies based on widely accepted assessment criteria.Therefore, we attempt to collect cumulative evidence of associations between genetic variants and EC risk from published meta‐analyses and GWAS, and evaluate these associations, which may offer referenced information for further investigation of genetic risk factors for EC and its subtype.22.1The following items were used in search process:  and (“cancer” or “adenocarcinoma” or “carcinoma” or “tumor” or “squamous cell carcinoma”) and  and (“genetic association” or “Genetic” or “SNP” or “polymorphism” or “single nucleotide polymorphism” or “genotype” or “variant” or “variation” or “mutation” or “susceptibility”). Additionally, we also checked all the relevant references to find other potential meta‐analyses that could offer relevant data.‐8 of P value was considered statistically significant; (f) they were studied of EC incidence/susceptibility .Meta‐analysis articles met to the following criteria: (a) The publications were in English; (b) cancer type was EC or including subtypes; (c) the patients with EC were diagnosed by pathological or histological examination; (d) sample size was not fewer than 1000; (e) they were studied of EC incidence/susceptibility . The following criteria should be met for screening the SNPs in GWAS on PubMed: (a) The publications were in English; (b) cancer type was EC, which includes all the subtype of EC; (c) the patients with EC were diagnosed by pathological or histological examination; (d) the studies included two phases (discovery and replication); (e) OR and 95% CI were provided and the less than cutoff of 1 × 102.2P value. In meta‐analysis, the following data were collected: first author, publishing year, gene name, genetic variant, OR and 95% CI,the number of studies, the number of subjects (cases and controls), ethnicity, I‐square, test for heterogeneity (Q test) between studies,Q testP value used for between‐study test of heterogeneity (Q‐test) was 0.10. P value >0.10 represents little heterogeneity while P value <0.10 indicates the presence of heterogeneity. We attempted to extract information of EC, including ESCC and EADC; however, the results showed that almost all of the articles were studied for EC and ESCC, not for EADC. Therefore, we were unable to conduct assessment of EADC, and then, we just evaluated cumulative epidemiological evidence for EC and ESCC. In addition, the eligible studies reported two major ethnicities, Asian and Caucasian. Majority of the meta‐analyses included a combination of two or more ethnicities, which were defined as “diverse populations.” Therefore, we extracted the information of diverse population, when applicable, Asian and Caucasian population were also extracted. Multiple studies concerning the same SNP reported conflicting results due to varied sample sizes and the selection of different association models. Therefore, given study quality and result credibility, we selected the most recently published study with the greatest number of and most integrated participants and the standardized report of the genetic association study based on guidelines of the Human Genome Epidemiology Network for systematic review of genetic association studiesP value was <0.05.Data were extracted by J.T and checked by two authors (C.L and GL). Information extracted from each eligible publication in GWAS included PMID of article, first author, publishing year, gene name, genetic variant, ethnicity of participants, the amount of subjects (cases and controls), minor allele frequency (MAF), OR, 95% CI, 2.3I2 < 25%), grade B (25% < I2 < 50%), or grade C (I2 > 50%). Protection from bias was mainly determined by sensitivity analysis and a series of bias tests including publication bias, small‐study bias, as well as an excess of significant findings  assay with a prior probability of 0.001 and an FPRP cutoff value of 0.2 to uncover potential false‐positive results among significant associations and evaluate whether these associations should be omitted, and we used the statistical power to detect an odds ratio of 1.5 for alleles with an elevated risk in FPRP calculations, as suggested by Wacholder et al.33.1As presented in Figure 3.2For 51 SNPs identified by meta‐analysis, 38 SNPs had statistically significant association with EC or ESCC susceptibility.CYP1A1 rs1048943, PLCE1 rs2274223, MMP2 rs243865 in EC and HOTAIR rs920778 in ESCC, from weak to moderate for ADH1B rs1229984 and COX‐2 rs20417 in EC, based on FPRP <0.05. Cumulative evidence were downgraded from strong to moderate for IL‐18 −607C>A, MMP1 rs1799750, SLC52A3 rs13042395 and MDM2 rs2279744 in EC, and TNF‐α rs1800629, C20orf54 rs13042395, microRNA124 rs531564 in ESCC, from moderate to weak for GSTT1 null/present, XRCC1 rs1799782, Hsa‐mir rs3746444, hOGG1 rs1052133, STK15 rs2273535 in EC, and microRNA‐34b/c rs4938723, NAT2 rapid/slow in ESCC, based on FPRP >0.2. Finally, five SNPs on five genes were rated as strong for cumulative epidemiological evidence of association by combining Venice criteria and FPRP results, including CYP1A1 rs1048943, EGF rs444903, MMP2 rs243865, PLCE1 rs2274223 for EC and HOTAIR rs920778 for ESCC. Seven SNPs with EC and three with EC were rated as moderate. Sixteen with EC and two with ESCC were rated as weak. Great discrepancy between the calculated amount and the true amount makes the grade arduous to determine. Therefore, calculated amounts of less than 3000 were not included for assessment in MAFs obtained from the dbSNP.As presented in Table C20orf54 and rs2274223 on PLCE1) were performed both in meta‐analysis and in GWAS.The associations including 17 SNPs identified in six GWAS studies were presented in Table 3.3XRCC1) with sample sizes >10 000 were also not significantly associated with EC; further investigations for this variant may not be fruitful. Certain variants presented with relatively small sample sizes; as such, the evidence for nonsignificant , Fas 1800682, XPA rs1800975, microRNA146 rs2910164, microRNA196 rs11614913, XRCC1 rs25487. The results of two variants, EPHX1 (Try113His and His139Arg)XRCC1 (Arg399Gln and Arg194Trp),Twenty variants presented inconsistent results in our review  which may combine with DNA to form adducts via a series of biochemical reactions. The ultimate carcinogens converted from the DNA adducts were considered to be associated with the development of EC.EGF, located on chromosome 4q25‐q27,MMP‐2 is a sort of zinc‐dependent endopeptidases, which can regulate various cell behaviors such as tumor initiation and growth by modulating cell proliferation, apoptosis and angiogenesis.PLCE1, located on chromosome 10q23, participate in cell growth, differentiation, gene expression and oncogenesis.HOTAIR, located on chromosome 12q13.13, might transform normal cells to malignant.IL‐8 −607C>A, MMP‐1 rs1799750, TNF‐α rs1800629, MicroRNA124 rs531564, SLC52A3 rs13042395 and MDM2 rs2279744 based on a high FPRP (> 0.2). The FPRP method considers the P value, prior probability, and statistical power of the test; as we calculated FPRP at prior probability of 0.001 and used the statistical power to detect an odds ratio of 1.5 for alleles with an elevated risk in FPRP calculations, certain otherwise significant associations may have been excluded. Previous studies using different prior probabilities have classified their results as more noteworthy. Further investigations on these six variants may be necessary to analyze their associations in greater depth. Additionally, SNP (rs13042395) on C20orf54 also showing moderate evidence of association with ESCC susceptibility, which downgraded from strong to moderate. This SNP was both verified in GWAS and assessed in meta‐analysis. The association in GWAS, however, may be more statistically significant and convincing when related meta‐analysis results are inconsistent or the association evidences are not strong by combination Venice criteria and FPRP method. The Venice criteria could assess multiple reasons of potential bias such as genotype error or misclassification and ethnicity stratification, which were difficult to perform in meta‐analysis. Therefore, results might be more convincing if different weights in the Venice Criteria were reset.There are six variants showing moderate evidence of association in our review, all of which downgraded from strong to moderate: ADH1B rs1229984 and COX‐2 rs20417) with risk of EC was upgraded from weak to moderate based on FPRP <0.05. Additional assessment of two variants were necessary, particularly the variant (COX‐2 rs20417) since sample size of study for this variant are relatively small .Cumulative evidence of two variants (Thirteen variants were found not to be significantly associated with EC risk, to include nine variants on seven genes and two mRNAs in a sample of approximately 4000 patients, at approximately 85% power to detect an OR of 1.15 under different model for a variant with MAF of 20%. The MAFs of the aforementioned seven variants were almost all over 0.3 despite sample sizes >4000. We can safely conclude, therefore, that these seven variants are unlikely to be associated with EC (see Table Certain limitations do apply to this report. Although we did a comprehensive literature search, some articles may have been missed. A variability in sample size presented among different studies; the smaller sizes may have impacted the credibility of the data. The evaluated data extracted from only one source which may be main cause of bias. Finally, only the susceptibility/incidence between genetic variants and EC risk were evaluated; however, other roles of genetic polymorphisms such as tumor progression, metastasis, drug resistance for EC were not be assessed due to lack of data or information. Despite the limitations of our method, we believe that our study, as an updated summary and evaluation of the existing literature reporting genetic predisposition of EC, is of value for further genetic studies.This review evaluated the cumulative epidemiological evidence of significant associations by combining the Venice criteria and FPRP results which identified five SNPs having strong evidence of true association. Collectively, our review provides referenced information for further investigation into genetic susceptibility of EC and ESCC.None declared. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file.


Previous studies have demonstrated that serum amyloid A (SAA) levels are correlated with the clinical outcomes of solid tumors. However, the available data have not been systematically evaluated. The objective of the present meta-analysis was to explore the prognostic value of SAA levels in solid tumors.Eligible studies were identified from the PubMed, EMBASE and Science Citation Index electronic databases. The clinical characteristics, disease/progression-free survival (DFS/PFS) and overall survival (OS) were extracted from the eligible studies. The pooled hazard ratios (HRs) and 95% confidence intervals (CIs) were calculated with Stata 12.0 software. We also performed subgroup, meta-regression and sensitivity analyses.In total, 12 eligible studies including 2749 patients were enrolled in the present meta-analysis. The pooled HRs with 95% CIs showed that elevated levels of SAA were significantly associated with poor OS  and DFS/PFS  in patients with solid tumors. Although publication bias was seem found in the studies with regard to OS, a further trim and fill analysis revealed that the adjusted HR was 3.02 (95% CI 1.96–4.63), which was close to the original HR. Subgroup analysis confirmed an elevated level of SAA as a strong prognostic marker in patients with solid tumors, regardless of tumor type, detection method, cut-off value, sample size, area and variance analyses.Our meta-analysis indicated that elevated levels of SAA might be an unfavorable prognostic marker for OS in patients with solid tumors. Despite the efforts of the scientific community, cancer is still a serious public health problem worldwide. Based on GLOBOCAN estimates, approximately 14.1 million new cancer cases and 8.2 million cancer-related deaths occurred in 2012 worldwide [Previous studies have proposed the concept that chronic inflammation promotes cancer development and progression , 4. It hMeta-analysis has shown that a high level of C-reactive protein (CRP), another acute-phase protein, is significantly associated with the poor prognosis of some cancers, including esophageal, colorectal and urological cancers, and it has already been reported to be a prognostic marker in relevant cancers –11. LateA comprehensive search strategy was employed to search PubMed, EMBASE and Science Citation Index up to February 2019 without applying a start date limit. The terms neoplasms, serum amyloid A protein, prognosis and cohort studies were used as medical subject headings (MeSH) and key words at the same time.Eligible studies in this meta-analysis met the following inclusion criteria: (1) patients were pathologically diagnosed with any type of solid tumors. (2) SAA as isolated from serum samples. (3) The study was designed as a cohort study. (4) The hazard ratios (HRs) with 95% confidence intervals (CIs) for survival outcomes were reported or could be calculated from the available data in the study.The exclusion criteria were as follows: (1) the study did not report the prognostic value of SAA in solid tumors. (2) The study had a small sample size, with fewer than 50 patients. (3) The study was a case report, letter, conference abstract, review or duplicate article. (4) The article was not written in English.Two independent reviewers separately assessed the quality of the studies using the Newcastle–Ottawa Scale (NOS), which addresses three aspects, namely, patient selection, study comparability and study endpoints. The maximum possible NOS score is 9. Studies that earned scores ≥ 5 were considered high quality, otherwise, they was considered low quality and removed. Any disagreement in the quality assessment of studies was settled by discussion.The following information was extracted: surname of the first author, publication year, country, tumor types, clinical stage, patient number, methods of SAA detection, cut-off value, duration of follow-up, outcome, and variance analysis. Furthermore, HRs and 95% CIs were obtained directly from the eligible studies or estimated using the method suggested by Tierney et al. . The aboQ test and the I2 statistic. Significant heterogeneity was defined as P  < 0.1 or I2 > 50%. The random effects model was chosen to investigate the pooled HR when significant heterogeneity existed. Otherwise, the fixed effects model was used. To assess whether the results were influenced by other factors, subgroup, meta-regression and sensitivity analyses were conducted. Publication bias was tested by the funnel plot and Begg’s and Egger’s tests. If publication bias was found, a trim and fill analysis was used to evaluate the number of missing studies and recalculate the pooled risk estimate with the addition of those missing studies [All statistical analyses in the present meta-analysis were conducted using Stata 12.0 software. Pooled HRs and 95% CIs were used to evaluate the prognostic value of SAA levels in solid tumors. When the pooled HR was greater than 1, we concluded that an elevated level of SAA was a negative prognostic factor for patients. The heterogeneity of the pooled results was measured using Cochran’s  studies .In total, 261 articles were initially collected by a systematic literature search of the PubMed, EMBASE and Science Citation Index electronic databases. By reading the titles and author details, 183 articles were excluded due to not involving SAA or being duplicate articles. The remaining 78 articles were further evaluated by inspecting the abstracts, and 60 articles were removed according to the exclusion criteria. In total, 18 studies were assessed by reading the full text, after which 6 studies were excluded because HRs could not be obtained from them. Finally, 12 eligible articles were included in this meta-analysis. The flow chart of the study selection process is shown in Fig. In total, 12 eligible studies with 2749 patients published between 2000 and 2017 were included in the present meta-analysis. Seven studies were conducted in Asian countries , and 5 were conducted in non-Asian countries . Nine different types of cancer were involved in the eligible studies, namely, renal cell carcinoma, breast cancer, gastric cancer, esophageal squamous cell carcinoma, rectal cancer, hepatocellular carcinoma, pancreatic cancer, nasopharyngeal carcinoma and non-small cell lung. The level of SAA in these studies was mostly measured using the nephelometry method, while enzyme-linked immunosorbent assays (ELISA) and latex agglutination turbidimetric immunoassays (LTIA) were also utilized. The cut-off value for the level of SAA was reported in 11 studies, and it ranged from 19.2 ng/ml to 22.0 mg/l. Among the included studies, OS was reported in 12 studies, and DFS/PFS was reported in 3 studies. The results of the studies were analysed by multivariate or univariate methods. Furthermore, only one study required extrapolation of the HR for OS, and the HRs and 95% CIs for both OS and DFS/PFS were directly reported in the other studies. The quality assessment scores of the included studies ranged from 5 to 9, all of which were regarded as high quality . The pooled HR for OS was 3.01 , which suggested that an elevated level of SAA was significantly associated with poor OS in cancer patients . The pooled HR for DFS/PFS was 1.67 , which also suggested that an elevated level of SAA was significantly associated with poor DFS/PFS in cancer patients , but Begg’s test was significant (P = 0.047). Based on the trim and fill analysis for OS, no missing studies were imputed in the contour-enhanced funnel plots , indicating that cut-off value might help explain the sources of the inter-study heterogeneity. All results of the subgroup and meta-regression analyses for OS are shown in Table For the OS, subgroup and meta-regression analyses were performed according to tumor types, detection methods, cut-off values, sample sizes, areas, and variance analyses. In the analysis stratified by tumor type, elevated levels of SAA were significantly correlated with poor OS in urinary cancers , digestive system cancers  and other cancers . When different detection methods were considered, elevated levels of SAA were significantly correlated with poor OS in the LTIA group , the nephelometry group  and the ELISA group . In the subgroup analysis by cut-off value, we found that poor OS was significantly correlated with a cut-off value less than or equal to the median value of 5.3 mg/l  and greater than 5.3 mg/l . A similar result was also observed in the subgroup analysis by sample size, both sample sizes less than or equal to a median value of 143  and less than 143  were significantly associated with OS. In addition, we also observed that elevated levels of SAA were associated with poor OS in studies conducted in Asian countries  and in non-Asian countries . Moreover, an elevated level of SAA was significantly associated with poor OS in multivariate analysis  and univariate analysis . More importantly, meta-regression analysis found that cut-off value had statistically significant in the inter-study heterogeneity (SAA is an acute-phase protein mainly produced by the liver under the regulation of inflammation-associated cytokines in the course of acute and chronic inflammatory processes. However, SAA is also synthesized in extrahepatic tissues, including primary and metastatic cancer cell lines , 28. PreTo the best of our knowledge, our study is the first meta-analysis assessing the prognostic role of SAA in various solid tumors. The meta-analysis included a total of 2749 patients with solid tumors, including renal cell carcinoma, breast cancer, gastric cancer, esophageal squamous cell carcinoma, rectal cancer, hepatocellular carcinoma, pancreatic cancer, nasopharyngeal carcinoma and non-small cell lung. In this meta-analysis, by estimating the pooled HR of the included studies, an elevated level of SAA was found to be significantly related to poor OS and DFS/PFS in patients with solid tumors. Although publication bias for OS was found in the meta-analysis, the results after adjustment by the trim and fill method were consistent with the original results. For the OS, further sensitivity analysis demonstrated that the result was not affected after excluding any single study. Finally, subgroup analysis between SAA levels and OS were performed, and an elevated level of SAA was still a negative marker for OS when the patients were stratified by tumor types, detection methods, cut-off values, sample sizes, areas, and variance analyses. Although inter-study heterogeneity was not significantly decreased in the subgroup analysis stratified by several factors, meta-regression analysis subsequently found that cut-off value maybe an important source of heterogeneity. Based on the above evidence, the results of our study are stable and reliable.When stratified by tumor type, there was a trend for the association of increased SAA levels with poor OS to be the most sensitive for solid tumors in the digestive system, as the pooled HR of the subgroup with digestive system cancer was the highest of the entire group. However, this finding was not reported by previous studies. A similar result was observed in the subgroup analysis by cut-off value, the relationship between increased SAA level and poor OS was more sensitive for the subgroup with > 5.3 mg/l SAA than the subgroup with ≤ 5.3 mg/l SAA. This indicates that a greater tumor burden may aggravate the inflammatory response. Compared with ELISA, nephelometry showed greater sensitivity in predicting OS in the subgroup analyses by detection methods. Certainly, there is a need to confirm our results with further clinical trials.However, some limitations in the current meta-analysis need to be acknowledged. First, only English language articles were included, while articles written in other languages were excluded. Second, one article could not directly provide a HR and its 95% CI. Thus, we extracted the HR and 95% CI through the procedure recommended by Tierney et al. , which mIn conclusion, the current meta-analysis showed that elevated levels of SAA are significantly associated with poor OS and DFS/PFS in patients with solid tumos. Furthermore, SAA might be used as a novel biomarker to predict OS of patients with solid tumors. In the future, larger-scale, multicentre and prospective studies are needed to validate our conclusions.


P = 0.0134). However, we did not observe any significant associations between the recurrent miscarriage risk and the number of miscarriages in different age groups. In conclusion, our study indicated that the rs9839776 CT genotype may contribute to an increased risk of recurrent miscarriage in the southern Chinese population and that rs9839776 may act as a prognostic biomarker in recurrent miscarriage patients. However, an experiment-based study with a larger sample size should be performed to confirm these results.Genetic susceptibility may be involved in the onset of recurrent miscarriage. Previous studies have shown that some genetic polymorphisms that regulate cell migration are associated with susceptibility to recurrent miscarriage. The SOX2 overlapping transcript (SOX2OT) may regulate the migration and invasion of multiple tumor cells and is related to susceptibility to various diseases. However, whether lncRNA SOX2OT polymorphisms are related to recurrent miscarriage susceptibility is unclear. Therefore, we investigated the relationship between the lncRNA SOX2OT rs9839776 C>T polymorphism and recurrent miscarriage susceptibility. We recruited 570 subjects with recurrent miscarriage and 578 healthy control subjects from a population in southern China and used the TaqMan method for genotyping. We found a significant association between the rs9839776 CT genotype in the SOX2OT gene and an increased risk for recurrent miscarriage (CT vs CC: adjusted OR = 1.357, 95%CI = 1.065 − 1.728,  Recurrent miscarriage is defined as the loss of two or more pregnancies with an unknown etiology before the 20th week of gestation with the same male partner , 2. AlthlncRNA MALAT1 and lncRNA CCAT2, have been associated with susceptibility to recurrent miscarriage [Long noncoding RNAs (lncRNAs) are defined as RNA transcript molecules (longer than 200 nucleotides) that are not translated into proteins , 9. Rececarriage . TherefoThe SOX2 overlapping transcript (SOX2OT) is a lncRNA that is located in the SOX2 gene . Recent This study was approved by the Medical Ethics Committee of Guangzhou Women and Children's Medical Center . All female subjects who participated in this study signed and approved a written informed consent form.The study population consisted of a total of 570 patients with recurrent miscarriage  and 578 pregnant female subjects with 2 or more normal pregnancies (no miscarriages) as controls.μL of peripheral blood leukocyte samples from all participants using the Blood DNA Isolation Kit  according to the manufacturer's instructions. Specific single nucleotide polymorphism  genotyping fluorescent probes were purchased from ABI . SNP genotyping was performed on ABI Q6 instruments  with 384-well plates according to the TaqMan real-time polymerase chain reaction protocol. PCR amplification was performed in a final volume of 5 μL, which included 0.04 μL of primers, 2.5 μL of 2x Master Mix , 1.46 μL of ddH2O, and 2.5 ng of DNA.Genomic DNA was extracted from 200 χ2 test. In addition, the independent effects of each polymorphism on recurrent miscarriage risk were assessed by logistic regression analysis. An odds ratio (OR) with a 95% confidence interval (CI) was used to determine the association between polymorphisms and recurrent miscarriage. A P value ≤ 0.05 was considered statistically significant.The data were entered into a database contained in SAS software . Hardy-Weinberg equilibrium (HWE) for the control group and clinical and demographic characteristics of the two groups were compared using the goodness-of-fit P = 0.7278). In addition, approximately 55.26% of the patients with recurrent miscarriage experienced two or three miscarriages, and more than 44.74% of the patients experienced four or more miscarriages in this study.We recruited 570 patients with recurrent miscarriage between 20 and 46 years of age and 578 healthy control subjects between 20 and 49 years of age . There wpHWE = 0.252). We found a significant correlation between the rs9839776 CT genotype and an increased risk of recurrent miscarriage .The independent association between the SOX2OT genotype distribution of the SNP rs9839776 C>T and recurrent miscarriage risk was evaluated. The distribution of genotypes for rs9839776 C>T is shown in A stratified analysis of subjects by age and the number of miscarriages further evaluated the effect of the SOX2OT rs9839776 C>T polymorphism in patients with recurrent miscarriage and control subjects (as shown in In this case-control study with 570 cases and 578 controls, we evaluated the association of the SOX2OT rs9839776 C>T polymorphism with the risk of recurrent miscarriage in a southern Chinese population. We found a significant association between the rs9839776 CT genotype in the SOX2OT gene and an increased risk of recurrent miscarriage. However, we did not observe any significant association with the recurrent miscarriage risk or the number of miscarriages in different age groups. To the best of our knowledge, this is the first study to investigate the relationship between the lncRNA SOX2OT rs9839776 C>T polymorphism and susceptibility to recurrent miscarriage.The SOX2OT gene, which is one of the major regulators of pluripotency, is located at the 3q26.3-q27 site of the human chromosome . StudiesResearch reports on SOX2OT gene polymorphisms are still few. rs9839776 is an intron variant in the SOX2OT gene , and manThis study found that the number of miscarriages and the age of the patients at the time of pregnancy are risk factors for miscarriages. The risk of miscarriage was observed to increase sharply after the age of 35 from 9.5% in subjects 20-24 years of age to 76% in subjects aged 45 years and older. In clinical practice, females ≥ 40 years of age are less likely to become pregnant, and the risk of miscarriage increases , 48. HowThe current research has some limitations. First, the analysis of miscarriages included the age of the subjects and the number of miscarriages, but due to a lack of information, other factors, such as alcohol consumption, smoking, and family history, which may affect the results of this study, were not considered in the stratified analysis. Second, the study was limited to the southern Chinese population and did not assess cases and controls of other ethnic groups. Considering the large difference in the incidence of miscarriages among different ethnic groups, the distribution and function of rs9839776 C>T in different populations require further investigation. Third, we did not analyze the expression level of SOX2OT. The relationship between the rs9839776 C>T polymorphism and its expression level requires further investigation.In conclusion, this case-control study confirmed that the lncRNA SOX2OT rs9839776 CT genotype is associated with an increased susceptibility to recurrent miscarriage, but its risk effect was not different among female subjects of all ages, and there was no difference in the number of miscarriages. A larger sample size study and experiments should be performed to confirm the role of the lncRNA SOX2OT polymorphism in the susceptibility to recurrent miscarriage.


S-transferase mu (GSTM1) polymorphism is associated with non-viral hepatic cirrhosis (HC). However, some studies showed different views. Therefore, in this paper, a meta-analysis was conducted to get a more comprehensive understanding of GSTM1 polymorphisms in non-viral HC susceptibility. The results showed that GSTM1 null was associated with the increased risk of non-viral HC . Subgroup analysis of cirrhosis type revealed that GSTM1 null was a prominent risk factor for alcoholic HC . Meanwhile, subgroup analysis of population indicated that the significant differences only existed in Asian population . In hospital-based studies, patients with GSTM1 null were more likely in risk of HC . Subgroup analysis using genotyping method showed a significant association between GSTM1 null genotype and HC occurrence in the studies employing the multiple PCR genotyping method . Based on the results of this analysis, it was concluded that GSTM1 null genotype could increase the susceptibility of non-viral hepatic cirrhosis. In addition, alcohol intake, Asian ethnicity, sample source from hospital and multiple PCR genotyping method may also influence the susceptibility of hepatic cirrhosis.It is reported that glutathione  Hepatic cirrhosis (HC) expresses in dysfunction of liver due to normal liver tissue are gradually replaced by necrotic hepatocytes, which is caused by different etiology . PatientHC are mainly caused by hepatitis virus infection , heavy alcohol consumption and exposure to some chemical substances –7. At prS-transferase (GST) gene family is involved in the biotransformation phase II of harmful substances and has an important function to protect the cellular [α , μ (mu), θ (theta), and ∏ (pi) classes [S-transferase mu (GSTM1) is one of the most widely expressed gene [Glutathione cellular , 12. The classes . Glutathsed gene . Null gesed gene . It was sed gene . Soto-Qused gene  reportedA lot of studies were performed to explore the association of GSTM1 gene polymorphism with cirrhosis risk. GSTM1 null genotype has been found to be related with non-viral hepatic cirrhosis in some studies –20. HoweS-transferase M1 (GSTM1)”. Two independent reviewers screened the relevant articles using standardized screening guide. The eligible articles were enrolled in this meta-analysis according to the inclusion and exclusion criteria.Literatures were searched to find all the related articles in Pubmed, Web of Science, Embase databases  using the keywords “polymorphism”, “cirrhosis”, “chronic liver disease”, “glutathione S-transferase M1 and the non-viral hepatic cirrhosis. (b) All enrolled studies must be the case–control studies. (c) Hepatic cirrhosis must be diagnosed on the basis of liver biopsy. (d) Published in English. (e) Studies with enough data to calculate odds ratios and corresponding 95% confidence intervals  were included.Studies meeting all of the following inclusion criteria are included: (a) included studies must be concentrated on the relationship between glutathione Exclusion criteria were as follows: (a) reviews, abstracts, letters, comments, family-based studies and single-case reports were excluded. (b)The articles with insufficient data or overlapped data were excluded.We evaluate the quality of eligible studies with a modified 0–10 point scale, which is the appropriate quality assessment for case–control study , 24. QuaTwo investigators (M. Liu and Y. Gu) extracted data independently. Any disagreement was settled by discussion. The extracted data included name of the first author, year of publication, country, ethnicity, number of cases and controls, genotyping method, control sources and genotype distribution in cases and controls.S-transferase M1 and the cirrhosis. Pooled ORs were calculated using random-effect model (M–H heterogeneity method) or fixed-effect model (Mantel and Haenszel method). I2 index and p value of the Chi-squared test were used to inspect the heterogeneity among the enrolled literature [p < 0.05 and/or I2 > 50%), the random-effect model was used to estimate Ors [Z test and p value of 0.05 were used to judge whether the differences of OR values had statistical significance. Sensitivity analysis was conducted to assess the influence of individual studies. Begg’s test was applied to evaluate the publication bias [This meta-analysis was performed using STATA software . Crude odds ratios (ORs) and corresponding 95% confidence intervals (CIs) were calculated to assess the strength of association between glutathione terature . If notamate Ors , on the mate Ors . Subgrouion bias .The complete searching procedure is shown in Fig. p = 0.154, I2 = 37.8%, Fig. p = 0.013, Fig. The between-study heterogeneity of all the five eligible studies was first analyzed and no significant heterogeneity was found . The shape of Begg’s funnel plot is shown in Fig. The result of Begg’s test showed that there was no obvious evidence of publication bias . In addition, results from subgroup analysis classified by ethnicity indicated that HC risk of Asians with GSTM1 null was increased in . But it was not applicable for non-Asians. Furthermore, subgroup analysis was also conducted on cirrhosis type. In alcoholic hepatic cirrhosis group, the results are consistent with that of all enrolled studies .To date, it is the first time to perform meta-analysis to reveal the association between GSTM1 polymorphism and non-viral HC susceptibility. Hepatic cirrhosis is affected by many factors. Among these risk factors, genetic factors have become a research focus now. Many studies showed that GSTs played a crucial role in the etiology of HC. GSTM1 is a common type of GSTs gene. Some researchers reported that GSTM1 polymorphism was associated with the increased risk of non-viral HC. However, other investigations suggested that there was no relationship between GSTM1 polymorphism and non-viral HC. After the combination of these data, results of this meta-analysis revealed that GSTM1 null was a risk factor for susceptibility of HC  and acetaldehyde, which are both associated with the developing of cirrhosis . HoweverThis meta-analysis is rigorous. First, this paper is focused on GSTM1 polymorphism and the risk of non-viral hepatic cirrhosis. Studies were selected from three open classic biomedical databases, Pubmed, Web of Science and Embase database. A thorough search strategy was designed. Language type and the period covered by the publications were also limited strictly. Second, objective quality evaluation, particular inclusion criteria and strict exclusion criteria were established to ensure the reliability of this meta-analysis. Finally, the results were generated through appropriate statistics. Sensitivity analysis and stratification analysis were also performed to control the confounding factors.However, this meta-analysis has limitations to some extent. The data of this meta-analysis were collected from published literature and it is impossible to eliminate publication bias completely. We can only minimize the effect of publication bias to obtain more reliable results. We studied both alcoholic and non-alcoholic HC in the subgroup analysis. However, most of the studies enrolled participants are the excessive drinker. Thus, it is hard to get an accurate result of the non-alcoholic HC in this work. Further well-designed studies focusing on non-alcoholic population with larger sample sizes and different ethnic population are needed to clarify the present findings.According to the analysis results of this study, GSTM1 null is associated with the increased risk of non-viral hepatic cirrhosis. Subgroup analysis of cirrhosis type, population, controlled source and detection method also suggest that GSTM1null is a prominent risk factor of hepatic cirrhosis. Thus, GSTM1 polymorphism is related to the pathologies of non-viral hepatic cirrhosis.


Previous studies found inconsistent results on the relationship between post-traumatic stress disorder (PTSD) and concentrations of 24-hour (24-h) urinary cortisol. This study performed a systematic review and meta-analysis to summarize previous findings on this relationship.We searched in the databases of Web of Science, PubMed, Embase, and Psyc-ARTICLES for articles published before September 2018. We used the random-effects model with restricted maximum-likelihood estimator to synthesize the effect sizes by calculating the standardized mean difference (SMD) and assessing its significance.p = 0.02). Subgroup analysis revealed that the concentrations of 24-h urinary cortisol were lower in PTSD patients than in the controls for studies that included female participants or studies that included participants from the United States of America.Six hundred and nineteen articles were identified from the preceding databases and 20 of them were included in the meta-analysis. Lower concentrations of 24-h urinary cortisol were observed in patients with PTSD when compared with the controls (SMD = -0.49, 95%CI , Overall, decreased levels of 24-h urinary cortisol were linked with the pathophysiology of PTSD. Nonetheless, more studies should be conducted to validate the molecular underpinnings of urine cortisol degeneration in PTSD. As a complex multifactor psychiatric disorder, post-traumatic stress disorder (PTSD) is triggered by unusual shocks or catastrophic events , and itsAlong with genetic predisposition, and environmental factors , PTSD coThere are various biological specimens such as plasma, serum, saliva, hair, cerebrospinal fluid and urine that could be used to measure cortisol –21. NeveIn future, it is expected that 24-h urinary cortisol could be a quick biomarker assay to assist in screening patients for PTSD. Studies on 24-h urinary cortisol alterations in screening PTSD have so far yielded inconsistent results both in direction and magnitude of hormonal changes. Although most studies found that PTSD was associated with lower 24-h urinary cortisol output, some earlier studies showed contrary results ,27.There had been no meta-analysis primarily examining 24-h urinary cortisol as a biomarker for PTSD screening . HoweverThe comprehensive literature search for relevant studies followed the guidance by the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) criteria (last search updated in September 2018) . This syThe search for relevant articles was restricted to articles published in the English language. This was performed in the following four electronic databases: Web of Science, PubMed, Embase, and Psyc-ARTICLES. The search strategy was designed in consultation with experienced librarians. For example, a search strategy in the database of Embase was structured as follows using keywords (search terms):  and . In addition, hand searching was conducted in the lists of references of the retrieved articles by XP and AC. Any inconsistencies between them were resolved by group discussion and consensus with a third party AL.A study eligible for this meta-analysis had to meet the following inclusion criteria: (1) the study had to be a case control study, which included a control group and PTSD cases, (2) the study reported the mean and standard deviation (SD) of the 24-h urinary cortisol levels, or these could be provided by the authors upon request, (3) the study assessed PTSD cases simultaneously with the assessment of the 24-h urinary cortisol levels, (4) the study was published in English, and (5) the study reported the diagnostic criteria for PTSD. In addition, studies were excluded if they (1) were review articles or case reports; (2) studied PTSD in combination with HPA axis disorder disease, or studied PTSD in patients with other mental illnesses, who used psychotropic medication or other medications which could influence the HPA axis and cortisol concentrations, and (3) studied non-humans or were vitro experiments or animal research.Two researchers [XW and ZW] independently screened and selected eligible articles . A thirdIn relation to the purpose of this study, the following information were extracted from the eligible studies by two independent investigators [AC and XW]: (1) name of first author, and publication year; (2) geographical area of the study; (3) characteristics of PTSD participants such as trauma type, age , gender distribution, and body-mass index (BMI) ; (4) sample characteristics such as sample size, and concentrations of 24-h urinary cortisol ; (5) PTSD assessment method; and (6) 24-h urinary cortisol collection and assay methods, intra-assay variation, inter-assay variation, storage temperatures and sensitivity. All the extracted data were organized in EpiData 3.0 and saved in Excel.The Newcastle-Ottawa Quality Assessment Scale (NOS) was used to assess the quality of the eligible studies . Therefo2 statistic  [All analyses were conducted in R software (version R i386 3.4.2). Accordingly, meta-analysis was carried out in the package, meta, and meta-regression analysis was performed in the package, metafor. The standardized mean difference (SMD) of the 24-h urinary cortisol levels between the PTSD and the control groups was calculated using Cohen’s d . Moreovegeneity) . In ordegeneity) .Finally, in all the statistical tests, the level of significance was set at the 5%, and all tests were two-sided. Furthermore, other sources of heterogeneity were explored using meta-regression analysis. Therefore, the following categorical variables were considered for the meta-regression model: country , trauma type , controls type , gender , PTSD assessment  = 1, other = 0), assayed method , frozen samples . In addition, the following continuous variables were considered for the meta-regression model: age, study quality and BMI.The search of literature yielded 619 relevant articles from Web of Science (373), PubMed (56), Embase (72), and PsycARTICLES (118). After that, 58 duplicates were deleted, leaving a total of 561 relevant articles. The abstracts of these articles were then reviewed to assess their eligibility. Following this assessment, 473 articles did not meet the inclusion criteria, hence they were excluded. Furthermore, full texts of the 88 articles were reviewed and this resulted in the exclusion of 68 articles. In the end, 20 eligible articles met the inclusion criteria and were included in the final meta-analysis .p = 0.021), but with considerable heterogeneity .p = 0.031), while for studies conducted elsewhere there was no significant difference. Additionally, subgroup analysis with respect to gender indicated that females with PTSD had lower concentrations of 24-h urinary cortisol than female controls , but no such difference was observed between males with PTSD and male controls. Besides, gender significantly explained a large amount of the overall heterogeneity, but still with residual heterogeneity (I2 = 11.3%). Furthermore, subgroup analysis by controls type showed significant lower concentrations of 24-h urinary cortisol in patients with PTSD than in the trauma exposed controls , while there was no significant difference between PTSD patients and the non-trauma-exposed controls. In the subgroup analysis according to assay methods, the concentrations of 24-h urinary cortisol were significantly lower in patients with PTSD than in the controls when radioimmunoassay (RIA) was used , while no difference was found when other assay methods were used.p = 0.0005), PTSD assessment  and age  in the meta-regression analysis model, the heterogeneous sources could be explained, and the difference was significant.There was little change in the SMD and corresponding 95% CI when studies were excluded one at a time, indicating low sensitivity of this meta-analysis. Also, no asymmetry was observed in the shape of the Egger’s funnel plot, and the p value (0.31) of the Egger’s test was not significant , implyinAccording to the best of our knowledge, this is the first meta-analysis comparing the 24-h urinary cortisol levels data between PTSD patients and the controls. In order to obtain relatively larger sample size, we expanded the scope of article searching in the online electronic databases, which brings 20 eligible studies with a total of 543 participants with PTSD and 738 controls. In the overall study sample, concentrations of 24-h urinary cortisol were lower in patients with PTSD than in controls. Therefore, the conclusions from this study are more comparative and persuasive compared with the previous studies. Besides, different from the preceding studies, this study only concerns with 24-h urinary cortisol concentration levels when studying its connection with PTSD.In addition to data on the differences in the 24-h urinary cortisol levels between PTSD patients and the controls, this study also analyzed almost all factors that may affect 24-hour urinary cortisol levels, including subgroup analyses. In particular, a lower 24-h urinary cortisol levels was found in PTSD patients as compared to the controls. However, the heterogeneity was quite large; suggesting that there may be differences based on some certain subgroups. It is worth noting that our study had significant heterogeneity among the studies, which is not surprising considering the differences between research variations, such as participant characteristics. The previous meta-analyses only included the 24-h urinary cortisol levels data in subgroup analysis, but no significant differences were found between PTSD patients and the controls because of the limited sample size of the studies used .Although previous studies had obtained consistent findings that plasma cortisol levels in PTSD patients was lower than that of the controls , they diAccordingly, the findings of this study suggested that females with PTSD had lower levels of 24-h urinary cortisol than female controls, whereas there was no significant difference of the same between males with PTSD and male controls. However, up to now, little is known about the biological mechanisms behind the preceding association between the 24-h urinary cortisol levels and PTSD in females. In spite of that, this association could explain why women are more vulnerable than men to the development of post-trauma symptoms, and take longer than men to recover from them . Also, tWhen 24-h urinary cortisol collection method , and stoAlso, age in this study explained some of the heterogeneity. This is consistent with results of a previous study, suggesting that the increased risk for developing PTSD is associated with lower concentrations of cortisol in both adults and children . That isHowever, this study has several limitations. First, due to the limited sample size, the effects of race, age and stressor patterns were not examined. Second, we were unable to use all studies in the subgroup analyses because not all eligible studies reported the subgroups of interest. Third, some studies did not report the use of urine standardization to control variability associated with urine dilution. Lastly, we included only articles published in English and excluded grey literatures. Therefore, this meta-analysis is prone to selection bias.Despite the preceding limitations, the results of this meta-analysis provide convincing evidence that lower concentrations of 24-h urinary cortisol may be associated with PTSD. However, it is worth noting that numerous often-overlooked factors may have a confounding influence on the concentrations of 24-h urinary cortisol. Therefore, future studies should elucidate whether low 24-h urinary cortisol is related to trauma type, age, or PTSD assessment methods.S1 TableDetails of search strategy.(DOC)Click here for additional data file.S2 Table(DOC)Click here for additional data file.


Kashin-Beck disease (KBD) is a serious human endemic chronic osteochondral disease. However, quantitative syntheses of X-ray detective rate studies for KBD are rare. We performed an initial systematic review and meta-analysis to assess the X-ray detective rate of KBD in China.For this systematic review and meta-analysis, we searched five databases , WanFang Data and the China Science and Technology Journal Database (VIP))using a comprehensive search strategy to identify studies of KBD X-ray detective rate in China that were published from database inception to January 13, 2018. The X-ray detective rate of KBD was determined via an analysis of published studies using a random effect meta-analysis with the proportions approach. Subgroup analysis and meta-regression were used to explore heterogeneity, and study quality was assessed using the risk of bias tool.p < 0.001). The pooled X-ray detective rate estimates were 11%  for northeast endemic areas, 13%  for northwest endemic areas, and 8%  for southwest endemic areas. There was a significant relationship between the survey year and the X-ray detective rate of KBD.A total of 53 studies involving 14,039 samples with X-ray detective rate in 163,340 observations in total were included in this meta-analysis. These studies were geographically diverse (3 endemic areas). The pooled overall X-ray detective rate for KBD was 11%  contains supplementary material, which is available to authorized users. Kashin-Beck disease (KBD) is a serious, endemic chronic osteochondral disease in humans that is distributed from northeast to southwest China, affecting 378 counties in 13 provinces , the preThe diagnosis and degrees of KBD is a crucial first step in the public health approach to confirm this type of osteochondrosis in China. X-ray image, considered as an important diagnostic criterion for KBD, could reflect the damage of articular cartilage before any clinical manifestations. For example, X-ray image can detect the lesions at epiphysis and metaphysis. However, the lack of consensus in diagnosing KBD has resulted in wide variations in reported X-ray detective rate. For example, according to reports of a national survey of KBD prevalence in 2005 . TherefoTo data, few studies have synthesized the results of KBD X-ray detective rate studies. Given the large number of X-ray detective rate studies that have been published in recent decades and the absence of nationwide quantitative estimates of the X-ray detective rate of KBD, it is an opportune time to perform a full systematic review and quantitative analysis of KBD X-ray detective rate.Here, we report the results of such a study in China to emphasize the need for more accurate nationwide and regional estimates of KBD X-ray detective rate. To analyze the wide variations in X-ray detective rate estimates, we investigated the influence of studies’ demographic and methodological characteristics.A comprehensive four steps search strategy was used to identify the relevant studies in this systematic review and meta-analysis. The study followed the Preferred Reporting Items for Systematic reviews and Meta-analysis (PRISMA) guidelines . We searData were extracted by three reviewers. The following data were extracted:1) study first author, 2) year and areas of publication, 3) number of KBD patients in the sample, 4) X-ray detective rate of KBD in the sample, and 5) age range of the sample. One reviewer crosschecked for accuracy. After removing duplicate studies, two reviewers screened the titles and abstracts using the eligibility criteria, with independent verification by other two reviewers. For suitable studies, we obtained the full text, and scrutinized the text against eligibility criteria, with independent verification by other two reviewers. The study quality was assessed by two reviewers using the standardized Risk of Bias Tool , 50%  and 100%  [P ≤ 0.05 indicated statistical significance.Heterogeneity was assessed by using the chi-square test of the Q statistic, which was quantified by the anation) . To explThe results of study identification, screening, eligibility, and inclusion are shown in the PRISMA flow diagram Fig. . We idenp < 0.001). Heterogeneity statistics  indicated that there was significant and substantive heterogeneity in the X-ray detective rate of KBD across studies  and Egger’s test  showed no potential risk of publication bias.The overall estimated X-ray detective rate of KBD was 11% . The estimated X-ray detective rate of KBD in the northwest was 13% . The estimated X-ray detective rate of KBD in the southwest was 8%   was significantly associated with the detected heterogeneity (Table P = 0.0001)  included in this study were retrospective observational studies, which are considered to provide moderate evidence. Thus, the conclusion drawn in this analysis is restricted by this study type. Finally, heterogeneity was observed in the study, which was not surprising as heterogeneity often exists in such meta-analyses of overall X-ray detective rate. Although subgroup and meta-regression analyses suggested that geographic region and survey year could explain a portion of the observed heterogeneity, the remaining heterogeneity among the studies was not explained by the examined variables.In conclusion, our meta-analysis found that the estimated the overall X-ray detective rate of KBD was 11% and ranged from 8 to 15% depending on the study. Further research is required to identify effective strategies for preventing and treating KBD.Additional file 1:Risk of Bias Tool: criteria for assessment of quality. (DOCX 13 kb)Additional file 2:Table S1. The correlation between the mean prevalence and potential sources. QS: quality score. (DOCX 14 kb)


Background: Thiazide diuretics may improve bone mineral density. However, results are inconsistent for studies evaluating the association between thiazides and risk of osteoporotic fracture. We performed an updated meta-analysis of cohort studies to determine the association between thiazides use and fracture risk.Methods: Relevant studies were identified via systematic search of PubMed and Embase. A random-effect model was used for meta-analysis. Subgroup analyses were performed to explore the potential influences of study characteristics on the outcome.Results: Seventeen cohort studies with 3,537,504 participants were included. The pooled results showed that use of thiazide diuretics at baseline did not significantly affect the risk of overall osteoporotic fracture incidence as compared with controls  with significant heterogeneity . Results of subgroup analyses indicated that general status of the participants may be an important determinant for the association between thiazide diuretics and subsequent risk of osteoporotic fracture. Use of thiazide diuretics was associated with significantly reduced risk of fracture in patients with acute status including new-onset stroke or spinal cord injury , but not in those with good conditions such as community-dwelling population or hypertensive patients (p for subgroup difference = 0.02).Conclusions: Use of thiazide diuretics is not associated with significantly affected risk of overall osteoporotic fracture. However, the association may be different according to the general status of the participants. Osteoporosis is a common skeletal disease characterized by increased bone fragility and risk of fracture . As an ivia their efficacy for lowering the urinary excretion of calcium   and CochDatabases of PubMed and Embase were searched from the index date for relevant records, using the combinations of the following terms (1) “diuretic” OR “diuretics” OR “thiazide” OR “thiazides” OR “hydrochlorothiazide” OR “chlorthalidone”, “bendroflumethiazide” OR “chlorothiazide” OR “cyclothiazide” OR “methyclothiazide” OR “hydroflumethiazide” OR “trichlormethiazide” OR “benzthiazide” OR “polythiazide” OR “buthiazide” OR “cyclopenthiazide” OR “metolazone” OR “quinethazone” OR “fenquizone” OR “clorexolone” OR “clopamide” OR “indapamide” OR “diapamide” OR “isoindapamide” OR “mefruside” OR “xipamide”; AND (2) “fracture”; AND (3) “prospective” OR “prospectively” OR “retrospective” OR “retrospectively” OR “followed” OR “follow-up” OR “cohort” OR “cohorts”. The search was limited to studies in humans and published in English language. The reference lists of original and review articles were also analyzed using a manual approach. The final literature search was performed on February 26, 2019.Articles were included in the meta-analysis if they met all of the following criteria: (1) published as full-length article in English; (2) reported as cohort studies ; (3) included adult population (≥ 18 years of age) without fracture at baseline; (4) use of thiazide diuretics were defined as exposure of interest at baseline, while participants that did not use thiazide diuretics were defined as controls; (5) documented the incidence of any osteoporotic fracture during follow-up; and (6) reported the risk RRs and their corresponding 95% CIs for the incidence of osteoporotic fracture comparing participants that used thiazide diuretics at baseline to those did not use thiazide diuretics. Reviews, letters, editorials, nonhuman studies, and studies with designs other than cohort study were excluded.Literature search, data extraction, and quality assessment of the included studies were performed independently by two reviewers (J.W. and K.S.) according to the predefined inclusion criteria. Discrepancies were resolved by consensus. Data that were extracted include: (1) name of first author, publication year, and country where the study was performed; (2) design characteristics (prospective or retrospective); (3) sources, characteristics, and numbers of the participants; (4) strategies to confirm the using of thiazide diuretics at baseline; (5) enrollment year and follow-up durations; (6) definitions of fracture outcomes, number of cases with osteoporotic fracture, and strategies to confirm facture outcome during follow-up; and (7) variables adjusted when presenting the results. The quality of each study was evaluated using the Newcastle-Ottawa Scale  which ra2 test were used to evaluate the heterogeneity among the included cohort studies  were calculated from 95% CIs or p values, and were logarithmically transformed to stabilize variance and normalized the distribution . The Coc studies . A signiogeneity . Sensitiogeneity . Predefiogeneity . We usedvia initial literature search of the PubMed and Embase databases, and two studies were found via manual search of the reference lists of the review articles. After exclusion of 152 duplications, 1,692 articles underwent screening. Subsequently, 1,629 were excluded through screening of the titles and abstracts mainly because they were not relevant to the purpose of the meta-analysis. Subsequently, 63 potential relevant records underwent full-text review. Of these, 46 were further excluded because 21 of them were case-control studies, 15 were reports of medications other than thiazide diuretics as exposure, four were repeated reports of included cohorts, four were studies evaluating the association between hyponatremia and fracture risk, and the other two did not report the incidence data of fracture outcome. Finally, 17 cohort studies were included . Results of sensitivity analyses by omitting one study at a time did not significantly change the results . Two of the studies included male participants exclusively (2 = 80%), but not in female participants  have significant influence on the association between use of thiazide diuretics and the risk of osteoporotic fracture.Our study has limitations which should be considered when interpreting the results. Firstly, significant heterogeneity exists among the included studies. Although we explored the potential source of heterogeneity by performing the subgroup analyses, these results should be interpreted with caution because limited numbers of studies were available for each stratum of subgroups. Moreover, we did not have access to individual-patient based data, and the subgroup analyses were performed on the basis of study-level results. The potential influence of study characteristics on the outcome should be confirmed in future large-scale cohort studies. Secondly, as inherited in the meta-analysis of observational studies, we could not exclude the possibility of other factors that may confound the association between use of thiazide diuretics and the risk of osteoporotic fracture despite that we pooled the result with the most adequately adjusted data. Thirdly, we were unable to evaluate whether the association between use of thiazide diuretics and the risk of osteoporotic fracture differed according to the doses and durations of thiazides treatment. Finally, whether the association between use of thiazide diuretics and the risk of osteoporotic fracture differed among individual medication of thiazide diuretics should also be determined.In conclusion, results of our meta-analysis indicated that use of thiazide diuretics is not associated with significantly affected risk of overall osteoporotic fracture. However, the association between use of thiazide diuretics and fracture risk may differ according to the general status of the participants.The raw data supporting the conclusions of this manuscript will be made available by the authors, without undue reservation, to any qualified researcher.JW conceived and designed the study. JW and KS selected the studies and collected the data. JW, LL, WS and SM analyzed data, and all authors interpreted the results. JW drafted and revised the paper. KS, LL, WS and SM revised the draft paper. All authors read and approved the final version of the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


Here, we carried out a meta-analysis to clarify polymorphisms between ApaI (C/A) (rs7975232), BsmI (G/A) (rs1544410), FokI (C/T) (rs10735810), TaqI (T/C) (rs731236) and Tru9I (G/A) (rs757343) in the VDR gene and PCOS susceptibility based on relative lager sample size.Polycystic ovary syndrome (PCOS) is a common endocrine disorder in reproductive-age women. Multiple susceptible gene as well as environmental factors and their interaction each other are contributed to the PCOS risk. Several case-control studies have researched the associations of the VDR gene variates and PCOS susceptibility published before 31th, May 2018. Crude odds ratios (ORs) and its 95% confidence intervals (95% CIs) in different comparisons were used to detected the strength of the association. All the statistical analyses of the present meta-analysis were performed by STATA version 12.0 software.English database of PubMed and Embase, and Chinese database of Wanfang and China National Knowledge Infrastructure (CNKI) databases were retrivaled for the relationship between VDR ApaI (rs7975232) polymorphism and PCOS susceptibility  was found in the overall population. After stratified by ethnicity, we showed that there is a significant association between VDR ApaI (rs7975232) polymorphism and susceptibility to PCOS in the Asian  population, but this association was not found in the Caucasian population. Additionally, a significant relationship between VDR BsmI (rs1544410) variates with PCOS susceptibility in the Asian  population, but this association was not found in the Caucasian population. We didn’t find any association between VDR FokI (rs2228570), VDR TaqI (rs731236), VDR Tru9I (rs757343) and PCOS susceptibility in the overall and the subgroup populations.Totally, 3587  participants from 13 studies were included which met our inclusion criteria. A statistically significant association between VDR ApaI (rs7975232) and VDR BsmI (rs1544410) polymorphisms are correlated with susceptibility to PCOS in the Asian population and VDR TaqI (rs731236), VDR FokI (rs2228570), VDR Tru9I (rs757343) did not reveal a relationship with the PCOS susceptibility.Our findings demonstrated that  A common endocrine syndrome, polycystic ovary syndrome (PCOS), is characterized by long-term absence of ovulation and high androgen, which is the most common causes of menstrual disorders and infertility in women during reproductive years , 2. The IR and hyperinsulinemia are frequent metabolic abnormalities in the PCOS, evidence suggests that vitamin D levels may be linked to hormonal and metabolic disorders . As a stVDR gene is located on chromosome 12cenq12 and contains 14 exons. Several single nucleotide polymorphisms (SNPs) in the VDR gene have been reported, such as ApaI in intron 8 (C/A) (rs7975232), BsmI in intron 8 (G/A) (rs1544410), FokI in exon 2 (C/T) (rs10735810), TaqI in exon 9 (T/C) (rs731236) and Tru9I in intron 8 (G/A) (rs757343). It has been shown that VDR polymorphisms  may contribute to the PCOS susceptibility, although the findings are as yet inconclusive [VDR gene polymorphism with incident PCOS outcomes, but only six studies were included for analysis [VDR gene polymorphisms and PCOS susceptibility.The nclusive . A previanalysis . CurrentPubMed, EMBASE, Chinese Wanfang, China National Knowledge Infrastructure (CNKI) and other databases were searched. The retrieval period is from establishment to May 31, 2018. The search terms and keywords are as follows: “vitamin D receptor or VDR”, “polymorphisms or variants”, and “polycystic ovarian syndrome or PCOS”. References to retrieved papers were also manually searched for other potential studies not included in the database.VDR gene with PCOS; (2) case-control study design; (3) genotype distributions were available for both cases and controls to calculate an OR and its 95%CI; (4) The diagnosis of PCOS is based on the Rotterdam criteria and the National Institute of Child Health and Human Development criteria [If the study met the following criteria, it was included in the meta-analysis: (1) study on the association of criteria , 29. ExcThe quality of the study was assessed using the Newcastle-Ottawa scale. The scale is composed of three aspects: selection, comparability and exposure, with a maximum score of 9 . A totalTwo researchers independently and carefully extracted the available data from each eligible study. Information on all eligible studies is as follows: (1) surname of the fist author; (2) publication year; (3) country of origin; (4) the ethnicity of population; (5) sample size of cases and controls. Our research team addressed the differences through discussion.VDR gene polymorphisms and PCOS risk were calculated using odds ratios (ORs) and its 95% confience interval (95%CI). All analyses used the allelic, recessive, and dominant genetic models. The chi-square test based on Q statistics was used to analyze the inter-study heterogeneity, which was considered to be significant when p value < 0.10 [2 test. When I2 was less than 50%, heterogeneity was acceptable, and the fixed effect model of mantel-haenszel method was adopted. Otherwise, the random effect model based on DerSimonian and Laird methods is adopted.The effect sizes of the association between the e < 0.10 . HeterogP-value of two-sided < 0.05 was considered statistically significant.We used Begg funnel plot and Egger’s linear regression test to assess potential publication bias. The Egger’s linear regression test examines the asymmetry of funnel plot measured on the natural log scale . One-wayAs Fig. ApaI rs7975232 (G > T) polymorphism was examined in 8 case-control studies [BsmI rs1544410 (A > G) variant was examined in 8 studies [Taq1 rs731236 (T > C) variant was examined in 10 studies [FokI rs2228570 (C > T) variant was examined in 7 studies [VDR Tru9I rs757343 with PCOS susceptibilitywas examined in 3 studies  variant and PCOS susceptibility are shown in Table VDR gene ApaI rs7975232 (A > C) polymorphism with PCOS susceptibility in the allelic , recessive  and dominant  genetic models in the overall populations. Population subgroup analysis showed that there was a significant correlation between VDR ApaI rs7975232 (A > C) polymorphism and PCOS susceptibility in the Asian population   and PCOS susceptibility. Significant heterogeneity was found in some comparisons and results from the random-effects model using the DerSimonian-Laird method were used. We did not find a correlation between VDR BsmI rs1544410 (A > G) polymorphism and PCOS susceptibility in all genetic models of the general population. Subgroup analysis by ethnicity revealed a significant association between polymorphism of VDR BsmI rs1544410 (A > G) and susceptibility to polycystic ovary syndrome in the Asian population   polymorphisms and PCOS susceptibility. Significant heterogeneity was found in most comparisons, and random effects model results on DerSimonian-Laird method were used. We found no correlation between variation and PCOS susceptibility in the general population and in subgroups by ethnicity.In Table VDR FokI rs2228570 (C > T) variant with PCOS susceptibility was found in the overall population and sub-groups by ethnicity.Table VDR Tru9I rs757343 (G > A) variation and PCOS susceptibility. The subjects of all included studies were conducted in the Asian populations. The heterogeneity test demonstrated no significant heterogeneity exist in all studies and fixed effects model results on Mantel-Haenszel method were used. There is no association between the VDR Tru9I rs757343 (G > A) polymorphism and PCOS susceptibility was found in the Asian population.In Table Except for the polymorphism of VDR FokI rs2228570 (C > T) and the risk comparison of PCOS in all white genetic models, as these comparisons included only two studies, the Begg and Egger trials were carried out in all comparisons. Begg’s funnel plots were performed in all comparisons showed the shape was symmetrical, and the Egger’s linear regression analysis further indicated that there was no publication bias in the meta analysis Table , Fig. 3.Sensitive analysis was conducted to estimate if our results were substantially affected by the presence of any individual. Our results suggest that no single study has a significant effect on the merger effect Fig. .Fig. 4SeVDR gene variants are associated with serum insulin levels in women with PCOS [VDR-mediated insulin resistance by regulating the expression of target genes [VDR gene involved in insulin signaling pathway is considered as an important candidate gene for PCOS [Genetic factors have become increasingly important in the progression of PCOS. Previous studies have shown that ith PCOS . Serum 2et genes . VDR genfor PCOS . Howeverfor PCOS , 36. Subfor PCOS .The metafor PCOS –39.VDR polymorphisms and PCOS susceptibility from available databases. The results included a total of 13 articles based on 1922 PCOS patients and 1665 controls, showed that VDR ApaI (rs7975232) and VDR BsmI (rs1544410) polymorphisms are associated with PCOS susceptibility and VDR TaqI (rs731236), VDR FokI (rs2228570), VDR Tru9I (rs757343) did’t reveal a relationship with the PCOS susceptibility. The results were in accordance with previous studies and might provide a new biomarker in the etiology of PCOS [VDR polymorphisms and PCOS risk. Subgroup analysis by ethnicity showed that VDR ApaI (rs7975232) and VDR BsmI (rs1544410) polymorphisms were significantly correlated with PCOS susceptibility in the Asian population but not in the Caucasian population. The reason for this finding may be genetic disparities between the ethnic groups. Due to the process of natural selection, different groups might have some differences in the functional variants [In this meta-analysis, we summarized the existing data on the associations of  of PCOS , 18, 20.variants .VDR gene polymorphism may play a role in the pathogenesis of PCOS by affecting the insulin signaling pathway [VDR gene would explain the observed association. In addition, VDR gene polymorphism may play a role in the pathogenesis of PCOS by affecting the PTH-vitamin D axis [VDR gene polymorphism is associated with serum PTH and 25 (OH) D levels, and vitamin D-VDR complex inhibits the secretion and synthesis of PTH [Accordingly,  pathway . Howevern D axis . Consists of PTH . Simsek s of PTH . Next, hs of PTH .VDR gene polymorphism in TaqI (rs731236) for T allele was significant association with PCOS and didn’t find any association between VDR ApaI (rs7975232), VDR BsmI (rs1544410), VDR FokI (rs2228570), VDR Tru9I (rs757343) and PCOS susceptibility in the all included studies. Such inconsistent results may be due to different number of studies included in the meta-analysis, different sample sizes and different statistical abilities. In their study, they conducted a meta-analysis and included 5 (9 studies in the present meta) studies on VDR ApaI rs7975232 (G > T), 4 (8 studies in the present meta) studies on VDR BsmI rs1544410 (A > G), 6 (10 studies in the present meta) studies on VDR TaqI rs731236 (T > C), 5 (7 studies in the present meta) studies on VDR FokI rs2228570 (C > T) and didn’t include VDR Tru9I rs757343 (G > A) studies (3 in the present meta) for analysis the association with PCOS susceptibility. In addition, we conducted a subgroup analysis by ethnicity among Asian and Caucasian populations, which was not included in their meta-analysis.The heterogeneity was observed in some comparisons, but partially it was resolved by subgroup analysis based on ethnicity. Our research found that VDR can act as an influencing factor on PCOS. These SNP mutations can be used as risk factors to evaluate PCOS.The results of this meta-analysis were different from those of Han et al. ‘s previous meta-analysis , which sAlthough the present meta-analysis has the advantage of a relatively large sample size for a combined result, several limitations should be addressed in interpreting our results. Firstly, we included relevant articles published only in English and Chinese so that potential language bias may exist in this study. Second, most of the studies were conducted in Asian populations, and the small number of studies in the Caucasian subgroup analysis may have resulted in insufficient statistical ability to detect subtle relationships. Third, age, gender, genetic variation, environmental factors exposure and other risk factors may have an impact on the incidence of PCOS, but this study only considered gene polymorphism. The effects of gene-gene and gene-environment interaction on the occurrence and development of the disease need to be further studied.VDR ApaI (rs7975232) and VDR BsmI (rs1544410) polymorphisms are associated with PCOS susceptibility in the Asian population and VDR TaqI (rs731236), VDR FokI (rs2228570), VDR Tru9I (rs757343) did’t reveal a relationship with the PCOS susceptibility. These results might not be generalized to other ethnic populations. Further studies with more sample size and including other confounding factors are still needed in the future for a definitive conclusion.In summary, current meta-analysis provided statistical evidence that