It has been reported that the functional telomerase reverse transcriptase (TERT) rs2853669 polymorphism might contribute to different types of human cancer. However, the association of this mutation with cancer remains controversial. Here, we conducted a meta-analysis to characterize this relationship.A systematic search of studies on the association of TERT rs2853669 polymorphism with all types of cancer was conducted in PubMed, Embase and Cochrane Library. The summary odds ratios (ORs) and corresponding 95% confidence intervals (95% CIs) were used to pool the effect size in a fixed-effects model or a random-effects model where appropriate. A total of 13 articles and 15 case-control studies, including 9,157 cases and 11,073 controls, were included in this meta-analysis. Overall, the pooled results indicated that the rs2853669 polymorphism was significantly associated with increased cancer risk in a homozygote comparison model . In the stratified analyses, a significant increased cancer risk was observed in Asian, but not Caucasian patients. A subgroup analysis by cancer type also revealed a significant increase in the risk of lung cancer, but not breast cancer.The results of this meta-analysis suggest that the TERT rs2853669 polymorphism is associated with a significantly increased risk of cancer, particularly lung cancer, in Asian populations. Telomeres are stretches of conserved, tandemly repeated DNA sequences that form the physical ends of eukaryotic chromosomes . TelomerRecently, many genome-wide association (GWA) studies focusing on cancer risk have been performed. TERT-locus single nucleotide polymorphisms (SNPs), located on chromosome 5p15.33, have reportedly been associated with the risk of several types of human cancers , 10. A TThe association between TERT rs2853669 and cancer risk has since been studied in multiple ethnicities and populations with inclusive results. Thus, we performed a meta-analysis screening of all relevant published data to clarify the association between the TERT rs2853669 polymorphism and cancer risk.Two independent investigators conducted a systematic published literature search in the PubMed, EMBASE, and Cochrane Library databases to identify studies evaluating the association between the TERT rs2853669 polymorphism and cancer risk conducted prior to October 14, 2016. The following keywords were used: “telomerase reverse transcriptase or TERT” and “cancer or tumor or carcinoma” and “polymorphism or polymorphisms or SNP or variants”. A search of all citations from the original studies was manually conducted to located further relevant studies.The inclusion criteria for studies in the current meta-analysis were as follows: (1) designed as a case-control study evaluating the association between the TERT rs2853669 polymorphism and cancer risk; (2) studied human tissues rather than animal tissues; (3) established definitive cancer diagnosis; and (4) contained available genotype frequency information to estimate the odds ratios (ORs) and 95% confidence intervals (95% CIs). The exclusion criteria were as follows: (1) duplication of same publications; (2) reviews, letters, posters and editorials; and (3) studies lacking detailed genotype frequencies.Two investigators independently extracted the usable data from the eligible studies, and a consensus was reached. In the case of a controversy, a consensus was reached through discussion. If the controversy remained, a third author would adjudicate the disagreements.http://ihg.gsf.de/cgi-bin/hw/hwa1.p) was used.The following data were sought from each article: first author’s name, year of publication, country of origin, ethnicity , cancer type, source of control , number of cases and controls with CC, CT and CC genotypes, and Hardy Weinberg equilibrium. When Hardy Weinberg equilibrium (HWE) in the control genotype was not calculated, an online program  was applied when P<0.10 and/or I2 index>50%. Otherwise, if the heterogeneity test value was P>0.10 and/or I2 index<50%, the fixed-effects model (Mantel and Haenszel method) was used. To evaluate the effect of each study on the combined ORs, a sensitivity analysis was conducted based on the leave-one-out method of each study in total and subgroups. In addition, subgroup analyses were stratified by cancer type , control type , and ethnicity . Eventually, Begg's [The heterogeneity in each study was evaluated based on Cochran’s Q statistic and the I, Begg's  funnel p, Begg's  tests weA total of 925 articles were acquired from the databases , and thirteen articles met the inclusion criteria –28. The According to all eligible studies, the results indicated a statistically significant association between the TERT rs2853669 polymorphism and cancer susceptibility in the homozygote comparison , homozygote comparison  and recessive model . However, no significant difference was observed in breast cancer in any genetic model , homozygote comparison  and recessive model , but not for Caucasians in any of the genetic models. For source of control, however, only one homozygote comparison  in the population-based control source showed a significant association between the TERT rs2853669 polymorphism and cancer risk.heterogeneity = 0.007) and dominant model . However, when stratified based on subgroup analysis, heterogeneity disappeared in breast cancer, which was stratified into cancer type and hospital-based control source subgroups.As shown in heterogeneity = 0.02) contributed to the source of heterogeneity, but not cancer types (Pheterogeneity = 0.15) or source of control (Pheterogeneity = 0.23). Consequently, ethnicity explained 44.67% of the between studies variance.Meta-regression was also used to examine the source of heterogeneity based on cancer types, source of control and ethnicity in the variant heterozygote comparison (CC vs. TT). The results indicated that ethnicity .Begg = 0.921, CC versus TT, Egger = 0.652, CC versus TT). Similarly, these results still did not show publication bias under the other genetic models . Consequently, there was no the evidence of publication bias in our meta-analysis.Begg’s funnel plot and Egger’s test were used to estimate the publication bias of eligible studies. No publication bias for the association between TERT rs2853669 polymorphism and cancer susceptibility was manifested based on Begg’s funnel plot , homozygote model (CC vs. TT) and recessive model (CC vs. CT + TT), also demonstrated a strong association of increasing lung cancer susceptibility in Asian populations. Notably, no significant increasing risk of breast cancer was observed in our meta-analysis.As shown in our meta-analysis, significant statistical heterogeneity was observed. Meta-regression was also used to detect the origin of this heterogeneity, showing that the ethnicity was the source of heterogeneity. However, when the stratification was based on subgroup analysis, heterogeneity disappeared in breast cancer, which stratified in the cancer type subgroup and hospital-based source control subgroup. Therefore, there is a strong foundation to consider between the etiology of different tumors and source of control groups, which also contributed to the source of heterogeneity. The results of the sensitivity analysis were reliable and robust. Additionally, neither the shape of the funnel plots nor the publication bias results showed statistical significance in this meta-analysis.Surprisingly, during the subgroup analysis stratified by cancer type, no significant increase in the risk of breast cancer was observed in any genetic model. However, our result is contrary to that of Li et al. , who conIn addition, the potential limitations of our meta-analysis should also be addressed. First, insufficient published studies were included in this meta-analysis, and more individual studies were required to determine a precise conclusion. Second, the results of gene-to-environment interactions were not obtained because of a lack of relevant information. Third, studies NO.12 and NO.14 did not meet the expectations of Hardy-Weinberg equilibrium, as both studies were focused on breast cancer. Fourth, in our meta-analysis, one cancer type (lung cancer or breast cancer) was only focused on one ethnicity (Asian or Caucasian). Therefore, additional high-level studies on different ethnicities are still needed.In conclusion, this meta-analysis suggests that the TERT genetic polymorphism rs2853669 is associated with an increased risk of cancer, particularly for lung cancer among Asians, while there is no significant association with increased risk of breast cancer. However, more functional studies with additional subgroups should be conducted to validate our findings.S1 File(DOCX)Click here for additional data file.S2 File(DOCX)Click here for additional data file.


Epidemiological literature regarding the effect of polycystic ovary syndrome (PCOS) as a risk factor for non-alcoholic fatty liver disease (NAFLD) remains inconsistent. Furthermore, it remains debatable whether NAFLD is associated with PCOS as a consequence of shared risk factors or whether PCOS contributes to NAFLD in an independent fashion. Therefore, this meta-analysis was conducted.This meta-analysis was conducted in accordance with the guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA). Relevant studies published before May 2017 were identified and retrieved from PubMed and Web of Science databases. The data were extracted, and the pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) were calculated.A total of 17 studies were included into the present analysis. Compared to the control group, the risk of NAFLD in the PCOS group was higher . When stratified by BMI and geographic location, the results indicated that the frequency of NAFLD risk was significantly higher in obese subjects , non-obese subjects , subjects from Europe , subjects from the Asia-Pacific Region,  and subjects from America . In addition, PCOS patients with hyperandrogenism (HA) had a significantly higher risk of NAFLD, compared with controls . However, there was no association between PCOS patients without HA and higher risk of NAFLD . The results of this meta-analysis should be interpreted with caution due to the small number of observational studies and possible confounding factors.The meta-analysis results suggest that PCOS is significantly associated with high risk of NAFLD. Although this association was independent of obesity and geographic region, it might be correlated with HA. At present, a question that should be addressed is whether non-alcoholic fatty liver disease (NAFLD) is associated with polycystic ovary syndrome (PCOS) as a consequence of shared risk factors, or whether PCOS contributes to NAFLD in an independent fashion.In the present study, a meta-analysis was performed to shed some light into this issue.A total of 17 studies were included into the meta-analysis. PCOS was significantly associated with higher risk of NAFLD, with an approximately 2.3-fold higher chance. This association was independent of obesity and geographic region but might be correlated with hyperandrogenism.In conclusion, the screening for NAFLD is only needed for hyperandrogenic women with PCOS, and not for all women with PCOS.Non-alcoholic fatty liver disease (NAFLD) is an increasingly common form of chronic disorder in the Western world, and is characterized by fat accumulation in the liver, which is independent of significant alcohol consumption, the use of steatogenic medications, or hereditary disorders , 2. NAFLAnother disorder associated with IR is polycystic ovary syndrome (PCOS). PCOS is an exceptionally common endocrine disorder in premenopausal women of reproductive age. It is characterized by hyperandrogenism (HA), chronic anovulation and polycystic ovaries (PCO) through ovarian ultrasonography, when other etiologies are excluded , 11. It Therefore, an update meta-analysis was conducted to investigate the association between PCOS and the risk of NAFLD development through studies published until May 2017. In order to explore whether NAFLD is associated with PCOS as a consequence of shared risk factors, or whether PCOS contributes to NAFLD in an independent fashion, stratified analyses were performed. Furthermore, factors associated with the presence of NAFLD in PCOS women were identified, and evidence of characteristics that can help identify women who have high risk of NAFLD were provided.This meta-analysis was conducted in accordance with the guidelines of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . The resA comprehensive search was performed on PubMed and Web of Science databases for relevant published articles and abstracts on the association between PCOS and NAFLD published until May 2017. Merely studies performed in humans and those reported in the English language were considered. For PCOS, the following search terms were used: ¨PCOS, polycystic ovary syndrome, polycystic ovary¨. These terms were combined with the following terms for NAFLD: ¨NAFLD, Non-alcoholic fatty liver disease, NASH, non-alcoholic steatotic hepatitis, hepatic steatosis¨. For the association between PCOS and NAFLD risk, “risk, risk factor” was used.Two independent reviewers (R-X.S. and J.W.) completed the initial screening process using pre-defined inclusion criteria (as detailed below). Articles were included for further review when these were original studies with primary data, involved human subjects, and were related to the topic of interest. These search results were independently reviewed by two reviewers and were compared for discrepancies. A third reviewer (X-Y.Y.) was involved to resolve any discrepancies that may occur between the two reviewers. In addition, the reference list of selected studies was manually reviewed for additional relevant publications.Studies were included in the meta-analysis when the following criteria were met: case-control studies, cohort studies and cross-sectional studies; studies that compared the prevalence of NAFLD in women with PCOS with matched controls; studies that contained sufficient data for analyses; studies published in the English language. For articles with overlapping data of the same population source, only the largest report was included.Articles were excluded when any of the following criteria were met: (1) review articles, meta-analyses, letters, commentaries, or case reports; (2) duplicates or continued work of previous publications; (3) studies without complete data; (4) articles that lack a control group; (5) articles not published in the English language. Studies without complete data, lack a control group and have overlapping data of the same population source were the common characteristics of the excluded papers.Data extracted from each study included the following: name of the first author(s), year of publication, geographic region from which the study population was derived, diagnostic criteria for PCOS and NAFLD, and the sample size. The baseline information and data were independently extracted by the two primary reviewers (X-Y.W. and J.W.) using the same standard. The results were compared, and disagreements were resolved by consensus.The quality of the included studies was independently assessed by two primary investigators (X-Y.W. and J.W.) using the Newcastle-Ottawa Scale (NOS) . Each stQ-test and I2-test. Heterogeneity was considered statistically significant at P < 0.05 or I2 > 50%, and a random-effects model was applied.The association between the prevalence of NAFLD in women with PCOS and matched controls was calculated using odds ratios (ORs) and 95% confidence intervals (CIs). In addition to the overall comparisons, subgroup analyses were performed based on geographic region, BMI, diagnostic criteria for PCOS and NAFLD, and androgen levels from articles with available adequate data. Heterogeneity between studies was assessed by P < 0.05 was considered statistically significant.In the assessment for publication bias, Begg’s test  and EggeA flow chart depicting the criteria for selecting studies for the present meta-analysis is shown in Fig. I2 = 4.6%, P-heterogeneity = 0.40). Therefore, the fixed-effects model was used for the meta-analysis. Compared to the control group, the PCOS group had a significantly higher risk of NAFLD  . Furthermore, the trim-and-fill method developed by Duval and Tweedie was used to adjust for bias [Begg’s test and Egger’s test were used to assess for publication bias for the included studies. There was evidence of publication bias in the overall meta-analysis for risk of NAFLD  gene, thereby prolonging the half-life of very low-density lipoprotein (VLDL) and LDL. As a result, lipids accumulate in the liver, and hyperandrogenic women with PCOS become more prone to fatty liver . Second,There were several limitations in the present meta-analysis. First, although the collected publications were all eligible, the studies included in the meta-analysis were limited. Meanwhile, these limited number of studies precluded any meaningful subgroup analyses stratified by important confounding factors such as IR. Second, the method used to assess NAFLD in the majority of studies was ultrasonography, and it has been demonstrated that liver biopsy is the gold standard for the diagnosis of NAFLD. The advantages of ultrasonography include non-invasiveness, reproducibility, low cost, and satisfactory sensitivity and specificity for epidemiologic studies. Third, publication bias existed among the involved studies. However, the trim and fill method was subsequently performed. The OR after adjustment changed to 2.08, which was almost consistent with the unadjusted result. Thus, the present results are reliable. Fourth, it was found that PCOS without HA was not associated with increased NAFLD risk, when compared to controls. However, few of the included studies had high heterogeneity. Finally, the entire body of evidence was observational, which may be biased due to unmeasured confounders. Although some studies were adjusted for important confounding factors, other modifiable factors were not accounted for in all these studies, such as family history, differences in diet, and/or physical activity patterns. Therefore, residual confounders may be present, which may influence these results. Thus, these observed effects may be attributable to these factors, as opposed to PCOS alone. Conversely, it could not be excluded that confounding factors may mask a more influential factor, leading to the underestimation of the effects of PCOS. Thus, future prospective high-quality studies may be necessary to conclusively address this point.In summary, the results of the meta-analysis suggest that PCOS is significantly associated with high risk of NAFLD. This association was dependent of androgen levels, which is the main feature of PCOS, and was interrelated to IR. These findings have clinical implications for NAFLD screening in hyperandrogenic women with PCOS.


Taq Ipolymorphism with prostate cancer risk, however these results were controversial. In order to provide a relatively comprehensive description of this relationship, we conducted this meta-analysis by searching PubMed, Embase, and China National Knowledge Infrastructure. Finally, 36 studies with 8,423 cases and 8,887 controls were included. Taq I polymorphism was found to marginally increase the prostate cancer risk in recessive genetic model (tt/Tt vs. TT: Odds Ratio (OR) = 0.89, 95% Confidence Interval (CI) = 0.80–1.00, p = 0.05) and allele genetic model  in the overall analysis. Subgroup analyses showed that significant increased risk was found in Asians in homozygote model  and allele genetic model , and in the subgroup of population-based controls in all the genetic models. These results suggest that Taq Ipolymorphism might be a risk factor of prostate cancer risk, especially in Asians. It could be considered as a promising target to predict the prostate cancer risk for clinical practice.Numerous previous studies reported the association of Vitamin D receptor gene  Prostate cancer (PCa) is second-most commonly diagnosed malignancy in males, and thought to be one of the leading causes of cancer-related death around the world. In 2014, approximately 233,000 newly diagnosed cases and 30,000 PCa-related deaths was reported in America [VDR) [2D3) is the hormonally active form of vitamin D. It binds to VDR and forms a heterodimer complex, which subsequently binds to the vitamin D response element and reduces the transcription levels of many genes that stimulating the cell growth and differentiation [Low serum levels of vitamin D might be one of the risk factors for PCa . LaboratVDR) . 1,25-Dintiation , 8.VDR gene and PCa risk has been the focus of research attention [Taq I polymorphism (rs731236) is one of the most widely-studied SNPs. It is a synonymous mutation located in exon 9 of VDR gene [VDR gene [Taq I variation might increase the susceptibility of PCa [Taq I polymorphism on PCa risk. Numerous studies in favor of the association of Taq I polymorphisms and PCa risk [Recently, the relationship of several single nucleotide polymorphisms (SNPs) of ttention , 9. Taq VDR gene . This muVDR gene . Recentlhism rs7336 is oney of PCa . HoweverPCa risk –19. MeanPCa risk –22. The PCa risk , 23, 24 PCa risk , 25–27. We identified 288 potentially relevant studies following the retrieval strategy. Based on the inclusion criteria, 36 studies , 25–49 bTaq I polymorphism marginally increase the PCa risk in the overall populations carrying TT genotype or T allele genotype , but not in other comparison models    Table .Taq I polymorphism could significantly increase PCa risk in the subgroup of population-based controls when patients carrying TT genotype or T allele genotype in all the genetic models  .As shown in Figure p = 0.004, I2 = 44%) Tt vs. TT: p = 0.000, I2 = 60%; tt/Tt vs. TT: p = 0.000, I2 = 56%; t vs. TT/Tt: p = 0.000, I2 = 54%; and t vs. T allele: p = 0.000, I2 = 57%)  or allele-frequency model   database searches were carried out for all the eligible papers. The following search terms were included: “Taq I polymorphism with PCa risk; (d) distribution of genotype frequency for cases and controls could be obtained or calculated; (e) and received more than 4 points in the NOS, which was considered to be high quality; (f) the difference of baseline characters and clinical information was not significant between PCa patients and controls.Retrieved studies were deemed eligible provided that they met all of the following criteria: (a) studies on human beings; (b) in a case-control or nested case-control design; (c) investigated the relationship of The studies meeting the inclusion criteria were read carefully by two investigators independently (Yansheng Zhao and Xiaofeng Liu). We collected the following information: author, year, country, ethnicity, genotyping methods, source of controls, sample size, and genotype and allele frequencies. The subjects were divided into different subgroups: Asians, Africans and Caucasians for ethnicity; hospital-based, population-based, and Benign Prostate Hyperplasia (BPH) for the source of controls; TaqMan, PCR-RFLP and SNPlex for genotyping method. The clinical stages were categorized as localized group (Gleason < 7) and advanced group (Gleason ≥ 7). In order to reach consensus on all of the items, any disagreement was resolved by a third reviewer (Lei Wang).2-test based on the Q statistic was conducted to evaluate the heterogeneity. The between-study heterogeneity was considered to be significant when I2 > 50% and p < 0.05, and the random effects model was used to combine values from studies [A χ studies . OtherwiThe pooled odds ratios (ORs) together with its 95% confidence intervals (95% CIs) were calculated to evaluate the strength of the association. The statistical significance of ORs was determined with Z-test. To get a more reasonable result, five genetic models were adopted in our analysis: homozygote model (tt vs. TT), heterozygous model (Tt vs. TT), dominant model (tt vs. TT/Tt), recessive model (tt/Tt vs. TT) and allele genetic model (t vs. T).p value < 0.05. Moreover, HWE of controls was recalculated with the goodness-of-fit χ2-test, P values of > 0.05 was considered as significant equilibrium.To assess the potential publication bias, Begg's Funnel plot was generated based on the analysis result and database size. The more asymmetry the funnel plot looked, the more publication bias was introduced. Meanwhile, Egger's test was also performed for further investigation. For the Egger's test, the significance level was set as For each outcome, we also performed subgroup analyses according to ethnicity, source of controls, genotyping method and clinic stages. Sensitivity analysis was performed to assess the stability of pooled results.P values of < 0.05 was considered to be statistically significant.All analyses were performed using STATA package version 12.0 . Two-sided


P < 0.00001). When stratified by injury sites, modest but statistically significant association was found in Achilles tendon pathology (ATP), anterior cruciate ligament injuries (ACLI) and tennis elbow (TE). Subgroup-analysis by ethnicity suggested that TT genotype of rs12722 was associated with tendon and ligament injuries in Caucasians  but not in Asians . Our findings indicated that rs12722 of COL5A1 was positively associated with tendon and ligament injuries, especially in Caucasian subjects. Individuals with TT genotype were predisposed to higher risk of ATP, ACLI and TE.The rs12722 polymorphism in COL5A1 gene has been implicated in the etiology of musculoskeletal soft tissue injuries in several association studies with limited sample size and conflicting results. The purpose of the present systematic review and meta-analysis was to evaluate and synthesize the currently available data on the association between rs12722 and musculoskeletal soft tissue injuries. Five electronic databases including Pubmed, EMBASE, ISI Web of Science, CNKI and Wanfang were searched to identify relevant studies published before 15 May, 2017. Summary odds ratios (ORs) and corresponding 95% confidence intervals (95% CIs) were estimated using the RevMan 5.3 software. Nine studies comprising 1140 cases and 1410 healthy controls met the eligibility criteria. Recessive model was confirmed to be the optimum model (TT vs TC + CC). The results indicated that rs12722 SNP was significantly associated with musculoskeletal soft tissue injuries (OR 1.58, 95% CI 1.33, 1.89;  Tendons and ligaments, both representing dense connective tissues and having comparably structural and mechanical properties, are essential constructions for the musculoskeletal system. The major duty of tendons is to deliver force from muscles to skeletons, causing the connected joints in locomotion, whereas the general function of ligaments is to stabilize joints and guide joints through their normal range of motion , 2. TypiOther than the two commonly accepted extrinsic causations, intrinsic factors such as genetic factors in conjunction with the gender, biomechanical, anatomic, hormonal, neuromuscular factors are as well supposed to contribute greatly to these injuries –12. CollUp to now, several studies focusing on the rs12722 SNP and the susceptibility of tendon and ligament injuries in separate population have been reported, but the sample size of individual study is limited and the findings were inconsistent. Thus, we conducted this systematic review and meta-analysis to investigate more accurate association between rs12722 SNP and tendon and ligament injuries.A total of 272 potentially relevant articles were identified by the initial literature search, including 119 from Pubmed, 24 from EMBASE, 124 from ISI Web of Science, 2 from Wanfang and 3 from CNKI. After the removal of 76 duplicated records, 196 studies were subsequently screened for eligibility with abstracts and titles. 181 records were removed because of irrelevance. The full-text of fifteen potentially related studies were downloaded to confirm eligibility, four of which were excluded for irrelevance, two studies were excluded due to unavailable data. Finally, nine case-control studies on rs12722 in COL5A1 gene were deemed eligible for inclusion in this met-analysis. The literature screen process was presented in Figure Nine case-control studies , 20–27 cThe NOS for non-randomized controlled trials were used to assess the risk of bias in included studies. The results for risk of bias judgement were consistent between two reviewers. Except for Kim et al. , all theP < 0.00001) and OR3  turned out to be statistically significant, whereas the OR2  was not significant, suggesting that recessive model could be the most appropriate model for meta-analysis. Using a recessive model, data for TC and CC were collapsed and compared to TT group. Since there was no evidence for between-study heterogeneity , the combined ORs showed that subjects with TT genotype had a 58 percent higher risk of musculoskeletal soft tissue injuries than people who had TC/CC genotypes  , a modest but statistically significant association could be detected in tennis elbow , Achilles tendon pathology  and ACL injuries   but not in Asians . However, the conclusion regarding the association could not be drawn in Asians due to limited number of included studies.Although there was no evidence for heterogeneity across included studies, we performed subgroup-analysis according to injury sites to determine whether rs12722 affect different sites of musculoskeletal soft tissue injuries in different ways. The estimated OR between different injury site was similar  and Begg’s test  also suggested no obvious publication bias. The sensitivity analysis by removing each individual study suggested that the results of our present study were stable (detailed data not shown). In particular, since Mokone et al.’s study  OR Collagen Type V Alpha I OR Collagen Type V Alpha1 OR Collagen Type 5 Alpha 1 OR Collagen Type V α1 OR Collagen Type 5 α1 OR Type 5 Collagen α1 OR Type V Collagen α1 OR Collagen 5a1 OR COL5A1) and  and . For Chinese academic databanks, we used “ji yin duo tai xing” and “COL5A1” to increase the sensitivity of our literature search. The bibliographic lists of relevant reviews were also manually searched for additional related studies.Studies satisfying the following criteria were included: (1) enrolled cases should be patients diagnosed as musculoskeletal soft tissue injuries ; (2) control groups should consist subject without any history of ligament and tendon injuries; (3) observational studies (case control or cohort studies) on humans that analyzed polymorphism in COL5A1 gene; (4) included articles should report sufficient data to estimate an odds ratio (ORs) with 95% confidence interval (95% CI).Studies were excluded for following reasons: (1) animal studies, case report or case series, expert opinion; (2) studies failed to provide sufficient data to calculate OR and the associated 95% CI; (3) studies duplicated for retrieval. The most comprehensive study was selected if data reported were duplicated or had been reported for more than once.Two investigators (Z. Lv and S. Gao) screened each article independently and were blinded to the findings of the other reviewer. According to the predetermined inclusion criteria, two reviewers conducted a strict screening to identify qualified articles independently, and they extracted data from these eligible papers using a standardized data collection sheet, which included first author, year of the publication, country, ethnicity of enrolled subjects, diagnosis of cases, genotypes distribution of cases and controls, Hardy-Weinberg equilibrium (HWE). Discrepancies between the two reviewers were resolved through discussion until a consensus could be reached. The third review author (P. Cheng) was consulted if a consensus could not be reached.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). Two reviewers assessed the methodological qualities of included studies independently, the results of risk of bias judgement were compared afterwards.The Newcastle-Ottawa Scale (NOS)  for the Recessive model: if OR1 = OR3 ≠ 1 and OR2 = 1;Dominant model: if OR1 = OR2 ≠ 1 and OR3 = 1;Complete over-dominant model: if OR1 = 1, OR2 = 1/OR3 ≠ 1;Codominant model: if OR1 > OR2 > 1 and OR1 > OR3 > 1, or OR1 < OR2 < 1 and OR1 < OR3 < 1.The goodness of fit of HWE was assessed using the Chi-square test in control subjects of each included study. The association between rs12722 and musculoskeletal soft tissue injuries was indicated as OR and 95% CI. To determine the most appropriate genetic model for COL5A1 polymorphism in the risk of ligament and tendon injuries, we used a genetic model-free approach . In brieQ-statistical test and I2 test. The random-effect model and fixed-effect model were used for data combination in the presence  or absence of heterogeneity  respectively. In case of statistically significant heterogeneity across studies, subgroup-analysis by injury sites was performed to test whether rs12722 affects different sites of musculoskeletal soft tissue injuries in different ways, subgroup-analysis by ethnicity was performed to test whether there is an ethnicity-specific effect. Sensitivity analysis was conducted by removing each study at a time and reevaluating the resulting effect on pooled data. Egger’s regression test and Begg’s rank correlation test were employed to estimate the potential publication bias . Funnel plots were generated using RevMan 5.3 software .Once the best-matching genetic model was identified, three genotypes were collapsed into two groups to obtain the pooled results. The heterogeneity between studies was determined using the Achilles tendon pathology, ACL injuries and tennis elbow along with other musculoskeletal soft tissue injuries are major health concerns affecting different levels of athletes and physically active individuals, making the genetic studies in sport increasingly important. Based on the findings of our current study, the polymorphism rs12722 of COL5A1 was positively associated with tendon or ligament injuries, individuals with TT genotype were predisposed to higher risk of Achilles tendon pathology, ACL injuries and tennis elbow. Genetic risk factors should be taken into account when developing multifactorial models to understand the molecular mechanisms underlying musculoskeletal soft tissue injuries. A further understanding of the relationship between gene variants and sport-related or occupational related soft tissue injuries will help clinicians and athletes to optimize the training programs, and thus effectively reduce the overall risk of injury.


It has been reported that CD40 rs1883832 might be associated with immune-related diseases susceptibility. Owing to mixed and inconclusive results, we conducted a meta-analysis of case–control studies to summarize and clarify this association.A systematic search of studies on the association between CD40 rs1883832 and immune-related diseases susceptibility was conducted in databases. Odds ratios and 95% confidence intervals were used to pool the effect size. 40 articles were included in our meta-analysis.CD40 rs1883832 is associated with decreased risk of Graves’ disease, especially in Asian; CD40 rs1883832 is associated with increased risk of multiple sclerosis; CD40 -1C>T (rs1883832) is not associated with the susceptibility of Hashimoto's thyroiditis, systemic sclerosis or Asthma; there is insufficient data to fully confirm the association between CD40 rs1883832 and systemic lupus erythematosus (SLE), rheumatoid arthritis (RA), Behçet's disease (BD), myasthenia gravis (MG), Crohn's disease (CD), ulcerative colitis (UC), Sarcoidosis, Fuch uveitis syndrome (FUS), Vogt-Koyanagi-Harada syndrome (VKH), Kawasaki disease (KD), giant cell arteritis (GCA) or Immune thrombocytopenia (ITP). CD40, expressed by some immune and non-immune cells, is a tumor necrosis factor receptor that plays a critical role in adaptive immunity. CD40 involves in B cell proliferation, the generation of plasma cells, B cell memory and the development of immature dendritic cells. –3 CD40 iCD40 -1C>T (rs1883832) is a functional single nucleotide polymorphism (SNP) located at -1 from the start codon of CD40 gene. The rs1883832T decreased the translational efficiency of CD40 transcripts, resulting in less CD40 protein level .Association between CD40 rs1883832 and immune-related diseases susceptibility has been studied in several populations. Sample sizes in these studies are relatively small. In Graves’ disease and Hashimoto's thyroiditis, meta-analysis has been performed before, , 7 howevclinicaltrials.gov, CNKI , WanFangData (one China database) and CQVIP (one China database) databases were carried out by two independent investigators. The following terms were used: “CD40 OR TNFRSF5 OR Bp50” AND “rs1883832 OR C/T-1 OR C>T-1 OR -1C/T OR -1C>T OR C-1T OR Kozak OR 5′-untranslated region OR 5′-utr”, without any limitation applied. The last search update was performed on August 4, 2016. References of related studies and reviews were also manually searched for additional studies. GWAS were searched in Immunobase.A systematic search in PubMed, Embase, Cochrane Library, Studies selected in this meta-analysis must meet the following inclusion criteria: (1) evaluation of the association between CD40 rs1883832 and immune-related diseases; (2) case-control study; (3) studies focusing on tissues of human beings; (4) detailed genotype data could be acquired to calculate the odds ratios (ORs) and 95% confidence intervals (95%CIs). Exclusion criteria: (1) duplication of previous publications ; (2) comment, review and editorial; (3) study without detailed genotype data; (4) GWAS; (5) studies focusing on cell lines. Dissertation thesis were included in the analysis.Study selection was achieved by two investigators independently, according to the inclusion and exclusion criteria by screening the title, abstract and full-text. Any dispute was solved by discussion.Two investigators extracted data of the eligible studies independently. In the case of a conflict, an agreement was reached by discussion. If the dissent still existed, the third investigator would be involved to adjudicate the disagreements. Try to contact the author by email for detailed genotype data.The following contents were collected: first author's surname, year of publication, disease type, the characteristics of cases and controls, source of control groups, country of origin, the detective sample, ethnicity, genotyping method, Hardy-Weinberg equilibrium, number of cases and controls for each genotype.The qualities of included studies were evaluated independently by two investigators according to Newcastle-Ottawa Scale (NOS)  and the P < 0.05 was considered as a significant departure from HWE. OR and 95% CIs were calculated to evaluate the strength of the association between CD40 rs1883832 and immune-related diseases. Pooled ORs were obtained from combination of single studies by allelic comparison (T vs C), dominant model (CT+TT vs CC), recessive model (TT vs CC+CT), homozygote comparison (TT vs CC) and heterozygote comparison (CT vs CC), respectively. The statistical significant level was determined by Z-test with P value less than 0.05.Our meta-analysis was conducted according to the PRISMA checklists.  Hardy-WeQ-test and I2 index. [Q-test's P-value was less than 0.10 and/or I2 index was more than 50%, the random-effects model (DerSimonian and Laird method) was used; otherwise, the fixed-effects model (Mantel and Haenszel method) was conducted. [P < 0.05 and P ≥ 0.05). Potential publication bias was checked by Begg's funnel [P value of Begg's test (PB) less than 0.05, and the P value of Egger's test (PE) less than 0.05 was considered a significant publication bias. All statistical analyses were performed with Stata 12.0 software . A two-tailed P < 0.05 was considered significant except for specified conditions, where a certain P value was declared.Heterogeneity was evaluated by 2 index.  When Q-tnducted.  Sensitivs funnel  plots ans funnel  An asymmclinicaltrials.gov = 0, CNKI = 118, WanFangData = 79, CQVIP = 12, other sources  = 1 [A total of 388 articles were acquired from databases  = 1 ). The serch) = 1 , 64; 3 drch) = 1 –16; 4 imrch) = 1 –20; 2 mirch) = 1 , 22). Firch) = 1 –62 were Summary results of each genetic model were listed in Table Sensitivity analyses were performed in any comparison and any subgroup including more than two studies. When study Tomer Y , Heward Because of the limited number of included studies, sensitivity analyses could not be performed in RA (rheumatoid arthritis), BD (Behçet's disease) and PB subgroup of SLE.Other results showed stability in sensitivity analyses. (Supplementary data).P value of Begg's test (PB) and P value of Egger's test (PE) were evaluated in every genetic model overall and in MS, GD, GD's Asian subgroup, GD's Caucasian subgroup. Significant publication bias was found in heterozygote comparison (CT vs CC) overall (PB = 0.043), dominant model (CT+TT vs CC) overall (PB = 0.000), recessive model (TT vs CC+CT) overall (PE = 0.021), recessive model (TT vs CC+CT) of MS (PE = 0.033), homozygote comparison (TT vs CC) of GD's Asian subgroup (PE = 0.016), dominant model (CT+TT vs CC) of GD's Caucasian subgroup (PE = 0.044). (Supplementary data).Begg's funnel plot and Egger's test were used to assess the publication bias. Symmetry of funnel plot, In GD, 5006 cases/4537 controls were involved, and we found CD40 -1C>T was associated with decreased risk of GD in any genetic model and the results showed stability in sensitivity analyses and no publication bias. Similar stable results were shown in GD's Asian subgroup in any genetic model, but significant publication bias was found in homozygote comparison (TT vs CC). Similar stable results were shown in GD's Caucasian subgroup in heterozygote comparison (CT vs CC) and dominant model (CT+TT vs CC), but significant publication bias was found in dominant model (CT+TT vs CC). This may reveal the difference in ethnicity.P value: 1.19E-04) performed by IMSGC et al. [In MS, 3851 cases/4368 controls were involved, and we found CD40 -1C>T was associated with increased risk of MS in 4 genetic models and the results showed stability and no publication bias. Similar stable results were found in its HWE and PB subgroups in allelic comparison (T vs C) and dominant model (CT+TT vs CC), however, publication bias analyses could not be performed. Our results in MS was consistent with the GWAS , SSc (2605 cases/3138 controls) and Asthma (675 cases/333 controls), no association was found in any genetic model and the results showed stability in sensitivity analyses.P = 0.03) of Joo YB [P = 0.004) of Wu C [P = 0.017) and no correction was done. Thus no certain conclusion can be drawn about the association between CD40 -1C>T and SLE susceptibility, and more studies were needed.In SLE (905 cases/1307 controls), no evidence was found for the association between CD40 -1C>T and SLE susceptibility in any genetic model, however, the results in heterozygote comparison (CT vs CC) lacked stability. In three studies of SLE, increased risk of SLE was found in dominant model (CT+TT vs CC: OR (95%CI) = 1.27 , f Joo YB  (593 cas of Wu C  (205 cas of Wu C  (107 casP = 0.038) and homozygote comparison (TT vs CC: OR (95%CI) = 0.735 , P = 0.035) of García BM [P = 0.002) and homozygote comparison (TT vs CC: OR (95%CI) = 1.565 , P = 0.025) of Chen F [In RA, BD, MG, CD, UC, Sarcoidosis, FUS, VKH, KD, GCA and ITP, less than three studies were included so that sensitivity analyses could not be performed. In two studies of RA, decreased risk of RA was found in allelic comparison (T vs C: OR (95%CI) = 0.89 ,  0.735 0.3, 0.978,arcía BM   CD40 -1C>T (rs1883832) is associated with decreased risk of Graves’ disease (GD), especially in Asian; (2) CD40 -1C>T (rs1883832) is associated with increased risk of MS (multiple sclerosis); (3) CD40 -1C>T (rs1883832) is not associated with the susceptibility of Hashimoto's thyroiditis (HT), systemic sclerosis (SSc) or Asthma; (4) there is insufficient data to fully confirm the association between CD40 -1C>T (rs1883832) and SLE, RA, BD, MG, CD, UC, Sarcoidosis, FUS, VKH, KD, GCA or ITP, and the results should be interpreted with caution. Well-designed studies with larger sample size and more subgroups are required to validate the risk identified in the current meta-analysis.


P = 0.33) suggested no potential publication bias. According to our results, urine UCA1 has greater diagnostic value in diagnosing bladder cancer, however further research studies with more well-designed and large sample sizes are required to confirm our findings.Urine UCA1 has been reported as a potential novel diagnostic biomarker for bladder cancer in several studies, but their results are inconsistent. As a result of this, a diagnostic meta-analysis to assess the diagnostic performance of urine UCA1 in detecting bladder cancer was conducted. A systematic electronic and manual search was performed for relevant literatures through PubMed, Cochrane library, Chinese Wan Fang and the China National Knowledge Infrastructure (CNKI) databases up to December 30, 2016. The quality of the studies included in this meta-analysis was assessed using the Quality Assessment of Diagnostic Accuracy Studies (QUADAS-2) tool. All analyses were conducted using stata12.0 software. Six studies collectively included 578 bladder cancer patients and 562 controls met the eligible criteria. The overall diagnostic accuracy was measured by the following: sensitivity 0.81 (95% CI = 0.75-0.86), specificity 0.86 (95% CI = 0.73-0.93), positive likelihood ratio 5.85 (95% CI = 2.72-12.57), negative likelihood 0.22 (95% CI = 0.15-0.32), diagnostic odds ratio 27.01 (95% CI = 8.69-83.97), and area under the curve 0.88 (95% CI = 0.85-0.91). Meta-regression analysis suggested that ethnicity significantly accounted for the heterogeneity of sensitivity. Deeks’ funnel plot asymmetry test ( Bladder cancer (BC) is the second most common urogenital malignancy, with 50% recurrence rate and 15-40% growing into muscle invasive disease . Early dLong non-coding RNAs (lncRNAs) are a class of transcribed RNA molecules with more than 200 nucleotides in length and lack protein coding function –11. AccuDespite many studies have demonstrated the potential of urine UCA1 as a novel diagnostic marker for BC, the previous studies have been limited by relatively small sample size recruited in the individual studies, and no previous published meta-analysis have addressed this research question. Thus, we carried out this meta-analysis to review and assess the overall diagnostic test accuracy of UCA1 for BC diagnosis.As shown in the flow diagram Figure , 351 potThe results of the QUADAS-2 study quality assessment were shown in Figure I2 = 66%, 95% CI = 36.34%-95.67%) and specificity . Therefore, the random effects model was applied for the further analysis. Overall, the sensitivity and specificity for the pooled data were 0.81 (95% CI = 0.75-0.86) and 0.86 (95% CI = 0.73-0.93), respectively. In addition, the pooled PLR was 5.85 (95% CI = 2.72-12.57), the NLR was 0.22 (95% CI = 0.15-0.32), and the DOR was 27.01 (95% CI = 8.69-83.97) , also suggesting that there was no threshold effect.Differences in cut-off values lead to the threshold effect. When there is a threshold effect, an inverse correlation is demonstrated among the sensitivity and specificity, leading to a typical ‘shoulder arm’ of the ROC plane distribution. Spearman correlation analysis is also suggests a strong positive correlation. In the current study, the representation of the sensitivity against the specificity of each study shown in an ROC plane Figure , displayP = 0.33).In our meta-analysis, Deeks’ funnel plot asymmetry test was conducted to evaluate potential publication bias Figure . No signBladder cancer is one of the most common male genitourinary tumors –27. To dIn current meta-analysis, we conducted the first diagnostic meta-analysis to assess the accuracy of urine UCA1 as a biomarker for bladder cancer. An AUC of 0.88 , with pooled SEN of 0.81  and SPE of 0.86  showed that UCA1 in voided urine sediment may be a promising biomarker to discriminate bladder cancer patients from normal bladder. As a prevalence-independent indicator, DOR value could indicate the degree of the association between diagnostic results and disease. The pooled DOR of 27.01  suggested that the overall accuracy of UCA1 for the diagnosis of bladder cancer is credible.The likelihood ratio (LR), including PLR and NLR, could also reflect the diagnostic accuracy. When positive likelihood ratio > 10 or negative likelihood ratio < 0.1, the likelihood of diagnosis or exclusion of a disease increased significantly. Nevertheless, in our meta-analysis, a pooled PLR of 5.85  and NLR 0.22  indicated that patients with bladder cancer have an ~5.86-fold higher chance of testing positive using UCA1 compared with controls and 33% individuals with bladder cancer have an negative result.LR and post-test probabilities are correlation with clinicians, due to they provide information about the likelihood of a patient with a positive or negative test actually exhibiting bladder cancer. From the Fagan's Nomogram, we found that when a pre-test probability of 20% was specified, the post-test probability positivity would raise to 59% with a positive likelihood ratio of 6, and the post-test probability negativity would decreased to 5% with a negative likelihood ratio was 0.22. These outcomes suggest a stable value for UCA1 in the diagnosis of bladder cancer.P = 0.397), also suggesting that threshold effect is not the cause of heterogeneity. Then, we found that sample size and ethnicity may have partially led to such heterogeneity. Therefore, subgroup analyses were performed to evaluate the contribution of the factors above sources of potential heterogeneity in sensitivity and specificity was explored by univariate meta-regression analysis and subgroup analysis. Urine UCA1 has higher diagnostic accuracy for bladder cancer detection in non-Chinese persons than in Chinese persons, indicating that ethnicity may be one of the causes of heterogeneity. Due to there were insufficient eligible studies to fully elucidate the source of the heterogeneity, the possible reason for the heterogeneity need to be investigated in future studies.Heterogeneity is a potential obstacle when interpreting the results for meta-analysis, which should be seriously considered . AlthougThe current study has several limitations. First, despite extensive literature search were performed, the number of included studies and sample sizes were small, which may restrict our ability to evaluate the accuracy of urine UCA1. Second, we could not determine the ideal cut-off value for urine UCA1 test, due to different cut-off values were adopted in each study. Third, this meta-analysis was a retrospective analysis, which may limit the conclusion due to selection bias. Fourth, only articles published in English or Chinese were enrolled in our meta-analysis, which may cause inevitable bias.Despite these limitations, the present evidence suggests that urine UCA1 is potential to be a diagnosis biomarker for bladder cancer, due to this non-invasive method has good overall diagnostic performance. However, large-scale and comprehensive studies must be performed in the future to validate this finding.This meta-analysis as conducted under the diagnostic meta-analysis guidelines . StudiesThe inclusion criteria were listed as follows: (1) articles were association between UCA1 and bladder cancer; (2) a diagnostic standard of bladder cancer was included; (3) sufficient data  for calculating sensitivity and specificity; (4) studies should base on humans; (5) studies were published in English or Chinese. Exclusion criteria were as follows: (1) studies without usable or overlapping data; (2) reviews, letters, case report and conference abstracts; (3) insufficient data for calculating sensitivity and specificity.Two investigators (Xiangrong Cui and Xuan Jing) retrieved the eligible publications independently. Any disagreement between the two investigators was determined through a discussion with a third investigator. The following data were extracted: first author, year of publication, country, ethnicity of patients, sample size, type of specimen, true and false positive and negative.I2 and Q test were performed to evaluate heterogeneity in SEN and SPE among included studies. If the tests show a P < 0.1 or I2 > 50%, the existence of significant heterogeneity would be verified, and then the random-effect model was employed. If not, the fixed-effect model was more appropriate [For the diagnostic meta-analysis, the accuracy indicator include the pooled sensitivity (SEN), pooled specificity (SPE), positive likelihood ratio (PLR), negative likelihood ration (NLR), diagnostic odds ratio (DOR), and their 95% confidence interval (CI) were calculated using the random-effect model. PLR was on behalf of the odds of positive test results of bladder cancer patients, while NLR reflected the odds of positive results in those without bladder cancer. DOR was the outcome of the combination of PLR and NLR (DOR = PLR/NLR). Simultaneously, the summary receiver operator characteristic (SROC) curve was created and the under the SROC curve (AUC) was calculated . The anaropriate , 35. Subropriate . All sta


Recent studies have shown platelet to lymphocyte ratio (PLR) to be a potential inflammatory marker in cardiovascular diseases. We performed a meta-analysis to systematically evaluate the prognostic role of PLR in acute coronary syndrome (ACS). A comprehensive literature search up to May 18, 2016 was conducted from PUBMED, EMBASE and Web of science to identify related studies. The risk ratio (RR) with 95% confidence interval (CI) was extracted or calculated for effect estimates. Totally ten studies involving 8932 patients diagnosed with ACS were included in our research. We demonstrated that patients with higher PLR level had significantly higher risk of in-hospital adverse outcomes  and long-term adverse outcomes . Sensitivity analyses confirmed the stability of our results. We didn’t detect significant publication bias by Begg’s and Egger’s test (p > 0.05). In conclusion, our meta-analysis revealed that PLR is promising biomarker in predicting worse prognosis in ACS patients. The results should be validated by future large-scale, standard investigations. Acute coronary syndrome (ACS), characterized by unstable atherosclerotic lesions, is the main cause of death from coronary heart diseasesThe interaction of inflammation and thrombosis is the critical factor in the pathogenesis of ACS5et al. demonstrated that PLR was positively associated with the Global Registry of Acute Coronary Events (GRACE) risk score and can improve its the predictive power for long-term cardiovascular events in patients with ACSet al. reported that PLR was an independent predictor of in-hospital cardiovascular mortality in patients with ST-elevated acute myocardial infarctionet al. found that there was a significant association between high PLR levels and the adjusted risk of 6-month all-cause deaths in ST elevation myocardial infarction underwent primary coronary intervention, while with no difference in in-hospital mortality among patients with different PLR levelsRecently, the prognostic importance of PLR has been investigated by several studies. Zhou We conducted a comprehensive computer search through the following databases from their inception until May 18, 2016: PUBMED, EMBASE and Web of Science. No language restriction was applied. The following terms was used to identify related articles: ‘platelet’ in combination with ‘lymphocyte’, and in combination with ‘acute coronary syndrome’ or ‘myocardial infarction’ or ‘myocardial ischaemia’ or ‘unstable angina’. The reference lists of all included studies were also screened to check for relevant articles.The inclusion criteria were listed as follows: (1) All patients were diagnosed as ACS, including acute ST elevation myocardial infarction, acute non-ST elevation myocardial infarction and unstable anginaTwo investigators (Qianqian Liu and Wenzhang Li) were responsible for data extraction and quality assessment. The following information were obtained: title, journal, author, language, year of publication, study design, country, sample size, mean age, gender, diagnosis, outcomes, adjusted or unadjusted risk estimates. The Newcastle-Ottawa scale (NOS) was used to assess the methodological quality of included studies, which is composed of three aspects , with 9 being the highest score. We considered studies with a NOS score more than or equal to seven to be of high quality. Discrepancies were discussed by the two reviewers.et al.2 tests. Fixed effect model was used if the p > 0.10 and I2 < 50% otherwise the random effect model was adopted14P values were two-sided.We performed meta-analysis on association between PLR and ACS prognosis using RR estimates and 95% CIs on a natural logarithmic scale. Compared with crude RR, the adjusted RR were preferred when available. We transformed adjusted odds ratio (OR) to RR when necessary with methods provided by Zhang Totally ten studies involving 8932 patients diagnosed with ACS were included in our final meta-analyses 8910171818910171819202122p = 0.180), while obvious in long-term adverse outcomes (p < 0.001).Eight studies, including 6083 patients were evaluated for in-hospital adverse outcomes while six study, involving 6253 patients were assessed for long-term adverse outcomes. The between-study heterogeneity was insignificant in in-hospital adverse outcomes (p < 0.05). Sensitivity analysis was shown in Compared with lower PLR level, patients with higher PLR level had significantly higher risk of in-hospital adverse outcomes , see The pooled RRs of long-term adverse outcomes were 2.32 (95%CI: 1.64–3.28) for patients with higher PLR level, see p = 0.662 and p = 0.121).Begg’s funnel plot and Egger’s test were performed to evaluate the publication bias of the included studies. The funnel plot was symmetric for both in-hospital and long-term adverse outcomes  and 7. Bet al.242526202891022PLR, as a systematic inflammatory response indicator, was initially introduced into clinical practice to improve the prognosis prediction of oncologic disorders, such as periampullary cancer, esophageal cancer and gastric cancer The between-study heterogeneity was insignificant when analyzing in-hospital adverse outcomes, which indicated the reliability of our pooling results. However, there was obvious heterogeneity when pooling the data regarding PLR and long-term adverse outcomes. Then we did subgroup analyses to search for potential factors that lead to heterogeneity. Our subgroup analyses demonstrated that the risk effects of higher PLR level in ACS adverse outcomes were not significantly influenced by potential confounding factors, such as age, sample size, region, outcome and risk factors adjustment. It should be pointed out that the between-study heterogeneity disappeared during the subgroup analyses according to region. We postulate that region may partly explained the observed heterogeneity and it might be ascribed to the following factors. First, as genetic variants have already be demonstrated to be related to hematological traits based on a series of genetic association studies, we believe that the predictive significance of PLR may also be different because of ethnicitic genotypic diversity32The underlying mechanisms of the association between higher PLR and worse prognosis seem to be multifactorial. Firstly, higher platelet counts serve to be both a result and a precipitating factor of inflammatory response. It is reported that megakaryocyte could be stimulated by several inflammatory mediators and presented accelerated proliferation and platelet-production35363741424344et al. found that patients with higher PLR levels exhibited higher peak CK-MB, creatinine, Killip class, admission glucose, and anemia in their study4647et al. showed a positive association between GRACE risk score and PLRThe significance of PLR in ACS prognosis prediction can also be validated through its association with other predictors. Ayhan It is interesting that, in some cases, thrombocytopenia may also indicate increased risk of ischemic events, which seems to be a paradox to our results4949et al. reported that increased red blood cell count is independently associated with risk of cardiovascular atherosclerotic events in their study of 6 years follow-up5354We had to admit that although there is good reason for PLR to be used as a predictive markers in worse prognosis among ACS patients, our research didn’t provide direct evidence of independent association between them. Other blood cell counts parameters, which are accompanied by or interfere with the change of PLR, were also demonstrated to be associated with adverse cardiovascular outcomes. For example, elevated red blood cell count, which is usually accompanied with activation, adhesiveness, and aggregation of platelet, can bring about increased blood viscosity and then further promote the genesis of ischaemic heart disease2, which were adopted to detect between-study heterogeneity in our study, can be misleading in meta-analyses if the appropriate reference Q value is not consideredSome limitations in our meta-analysis should be addressed. First of all, most studies included in our research were conducted in European countries, which may weaken the representative of results. Secondly, some studies did not adjust for potential confounders, which may limit the accuracy of the risk estimates. Thirdly, the cut-off point of PLR in two studies was not described, which limit the application of our research. Finally, some statistical limitations in our research should also be pointed out. No matter we use Egger’s or Begg’s test, we had to mention that the power to detect publication bias is low with a small number of included studiesIn conclusion, our meta-analysis revealed that PLR is promising biomarker in predicting both in-hospital and long-term worse prognosis in ACS patients. The results should be validated by future large-scale, standard investigations.How to cite this article: Li, W. et al. Platelet to lymphocyte ratio in the prediction of adverse outcomes after acute coronary syndrome: a meta-analysis. Sci. Rep.7, 40426; doi: 10.1038/srep40426 (2017).Publisher's note: Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.


XRCC1,XRCC3 and XPD genes with hepatocellular carcinoma risk by pooling the results of 20 English articles. This meta‐analysis was conducted according to the PRISMA statement, and data collection was independently completed in duplicate. In overall analyses, the minor alleles of four variants, Arg280His , Thr241Met , Asp312Asn  and Lys751Gln , were associated with the significant risk for hepatocellular carcinoma. There were low probabilities of publication bias for all variants. Subgroup analyses revealed significant association of XRCC1 gene Arg399Gln with hepatocellular carcinoma in Chinese especially from south China , in larger studies  and in studies with population‐based controls . Taken together, our findings demonstrated that XPD gene Asp312Asn and XRCC1 gene Arg399Gln might be candidate susceptibility loci for hepatocellular carcinoma. Considering the ubiquity of genetic heterogeneity, further validation in a broad range of ethnic populations is warranted.Hepatocellular carcinoma is a complex polygenic disease. Despite the huge advances in genetic epidemiology, it still remains a challenge to unveil the genetic architecture of hepatocellular carcinoma. We, therefore, decided to meta‐analytically assess the association of six non‐synonymous coding variants from  Liver cancer is rapidly escalating to epidemic proportions worldwide, and it constitutes a major burden on individual and public health. Liver cancer is much more common in men than in women, and it is a common malignancy in Asia and Africa XRCC1 (gene ID: 7515), XRCC3 (gene ID: 7517) and XPD (gene ID: 2068), in susceptibility to hepatocellular carcinoma, by pooling the summarized data from 20 English articles from medical literature.DNA damage is widely recognized as a major primary cause of cancer, and without correct repair, such damage can give rise to genetic instability and eventually carcinogenesis XRCC1, XRCC3 or XPD genetic variant(s) with the risk for hepatocellular carcinoma was carried out through searching PubMed and Embase using the predefined key words: (‘hepatocellular’ or ‘liver cancer’ in Title) and (‘XRCC*’ or ‘XPD’ or ‘DNA damage repair’ or ‘DNA repair’ in Title/Abstract) and . The latest search checkpoint was on October 19, 2015. Each source of relevant articles was integrated into the EndNote X5 software  for reference management. In addition, the bibliographies of some major narrative reviews, original and meta‐analytical articles were also abstracted to make sure there was no missing hit. The conduct of this meta‐analysis was in agreement with the guidelines formulated in the Preferred Reporting Items for Systematic Reviews and Meta‐analyses (PRISMA) statement (Table S1) Identification of relevant articles investigating the association of XRCC1 gene: Arg194Trp, Arg280His, Arg399Gln; XRCC3 gene: Thr241Met; XPD gene: Asp312Asn, Lys751Gln) were examined in association with hepatocellular carcinoma risk, and the count of each genotype was provided in both hepatocellular carcinoma patients and controls. In case of unavailable genotype counts, the article was includable if odds ratio (OR) and its 95% confidence interval (95% CI) were provided. Articles were excluded if they were conference abstracts or posters due to insufficient information or if they were editorials, narrative reviews, meta‐analyses or lacking of control groups.Articles were included if one or more of the six variants  independently collected and cross‐checked all necessary data from each qualified article, with any disagreement solved by a discussion. All necessary data included surname of the first author, year of publication, country where study samples collected and their acquisition time, sources of patients and controls, genotyping method, matched status, sample sizes, the genotype counts of six examined variants between the two groups and some baseline characteristics, including age, gender, smoking, drinking, hepatitis B and C virus infection, and family history of cancer.I2) was used to quantify the magnitude of heterogeneity across studies, and the I2, which takes values from 0% to 100%, represents the percent of the observed variability that results from heterogeneity rather than chance. Practically, the I2 of greater than 50% was indicative of significant heterogeneity.Given the small number of mutant homozygous genotypes, the association of six examined variants with hepatocellular carcinoma risk was assessed under allelic and dominant models only. The random‐effects model using the DerSimonian and Laird method was employed to compute weighted OR and its 95% CI. The inconsistency index .n = 14) and the Indo‐Pakistani region (n = 4). There were, respectively, 6, 4, 13, 6, 4 and 6 studies for variants Arg194Trp, Arg280His, Arg399Gln, Thr241Met, Asp312Asn and Lys751Gln. The baseline characteristics of these 20 studies are presented in Table Using the predefined key words, we have identified a total of 78 potential relevant articles from PubMed and Embase databases. Through a rigorous process of selection, 20 English articles involving 20 independent studies (6449 hepatocellular carcinoma patients and 8263 controls) were qualified for the final analysis P: 1.37, 1.13–1.66, 0.001), Thr241Met , Asp312Asn  and Lys751Gln  only, and no indication of heterogeneity was observed for Thr241Met (I2 = 16.5%) and Asp312Asn (I2 = 0.0%) (Table I2 = 0.0%) and Asp312Asn (I2 = 0.0%) only. However, after the Bonferroni correction for multiple tests (P < 0.05/6), only Asp312Asn persisted significant under both genetic models .Using the trim‐and‐fill method, one study was reported to be missing for Arg280His under allelic model and for Asp312Asn under both models , and no significance was found in subjects from the Indo‐Pakistani region and France. Second, after grouping studies by total sample size at a cut‐off of 500, there was no significance in smaller studies under both genetic models, but in larger studies, the 399Gln allele was associated with the significant risk of hepatocellular carcinoma under both allelic  and dominant  models. Third, regarding source of controls, significance was only identified for Arg399Gln in studies recruiting controls from populations  rather than from hospitals. Fourth, upon stratification by genotyping method, Arg399Gln variant was not associated with hepatocellular carcinoma under both genetic models for studies using restriction fragment length polymorphism (RFLP) method and the others. Fifth, restricting analysis to studies with matched age between patients and controls identified significant association with hepatocellular carcinoma . Sixth, to exclude the confounding impact of disproportionate hepatitis B virus infection, we analysed this association in studies with the comparable percentage of hepatitis B virus infection between the two groups, and there was enhanced association under both allelic  and dominant  models.In view of the limited number of qualified studies, stratified analyses were only undertaken for Arg399Gln  and controls (P = 0.008) were found to be significant confounders for the association between Arg399Gln variant and hepatocellular carcinoma risk.Further in meta‐regression analysis, the percentages of hepatitis B virus infection in patients (XPD gene Asp312Asn was a candidate locus predisposing individuals to hepatocellular carcinoma, and for the widely evaluated variant Arg399Gln in XRCC1 gene, country, sample size, hepatitis B virus infection and source of controls might confound its association with hepatocellular carcinoma.The aim of this comprehensive meta‐analysis was to systematically assess the susceptible roles of six non‐synonymous coding variants in three DNA repair genes from the English literature. The key findings of this meta‐analysis are that XPD gene increased the risk of having hepatocellular carcinoma, irrespective of genetic models, and there was no indication of heterogeneity and publication bias. However, a note of caution should be sounded when interpreting this finding as only four studies were available for summarization, and moreover as we only analysed data in articles published in English language, the possible existence of publication bias cannot be excluded convincingly It is widely accepted that hepatocellular carcinoma is a multifactorial disease, and generally, carcinogenesis can be promoted by a single dominant mutation leading to expression of an oncogene. It is of importance to capture targeted genetic culprits responsible for the functional changes of these oncogenes. The interrogation of non‐synonymous variants in coding region is a promising choice as they might change protein's function and consequently tailor susceptibility to disease XRCC1 gene Arg399Gln and hepatocellular carcinoma was negative, further stratified analyses revealed that country, sample size, hepatitis B virus infection and source of controls constituted the potential sources of heterogeneity. It is of particular interest to notice that Arg399Gln variant was associated with the significant risk of hepatocellular carcinoma in Chinese subjects, especially from south China, suggesting the existence of genetic heterogeneity. This association was deemed robust as significance remained after restricting analysis to larger studies only. Moreover, it is widely recognized that the coinfection of hepatitis B virus is an established risk factor for hepatocellular carcinoma, and this infection was unlikely able to cloud the susceptibility of Arg399Gln variant to hepatocellular carcinoma. On the basis of these observations, our meta‐analytical findings suggested that XRCC1 gene Arg399Gln was another susceptibility locus for hepatocellular carcinoma in Chinese. However, considering the limited number of qualified studies, this result cannot be directly extrapolated to the other ethnic groups, and further investigations in groups of other continents are timely required.Although the overall association between Several possible limitations should be acknowledged in this meta‐analysis. First, only English articles were retrieved, and selection bias cannot be totally ruled out, although our trim‐and‐fill method revealed a low probability of publication bias. Second, only six non‐synonymous variants from three DNA repair genes were meta‐analysed, and other functional variants such as in the promoter regions of other relevant genes also deserved attention pending sufficient published data. Third, besides Arg399Gln, there were limited numbers of qualified studies for the other examined variants, which precluded further exploration on heterogeneity by using stratified and meta‐regression analyses. Fourth, all involved studies were retrospective in nature, and it is intriguing to see whether the two significant variants were associated with the relapse, metastasis and survival in patients with hepatocellular carcinoma following hepatectomy.XPD gene Asp312Asn and XRCC1 gene Arg399Gln might be candidate susceptibility loci for hepatocellular carcinoma. Considering the ubiquity of genetic heterogeneity and in view of small sample sizes involved, our findings should be considered to be preliminary until being replicated or confirmed in other larger, well‐designed studies in the future. For practical reasons, successful confirmation of our findings would facilitate the identification of individuals at high risk for developing hepatocellular carcinoma in future clinical screening.In conclusion, we through a comprehensive meta‐analysis demonstrated that Figure S1 The selection process of qualified articles in this meta‐analysis.Click here for additional data file.Table S1 The PRISMA checklist.Click here for additional data file.


In the context of supplementary antioxidants having no anticancer effect, it is important to update the meta-analysis to evaluate whether there is an association between intake of citrus fruit and gastric cancer risk.The list of articles to be searched was established using citation discovery tools provided by PubMed and Scopus. The effect size of each article to be used in meta-analysis was calculated using the interval-collapse method. Summary effect size (sES) and 95% confidence intervals (CI) were obtained by conducting this meta-analysis. Random effect dose–response meta-regression (DRMR) was performed to investigate the dose–response relationship.A total of five cohort studies were selected. The result was 13% reduction of gastric cancer according to the intake of citrus fruit . In subgroup analysis, it was found that the intake of citrus fruit inhibited cardia gastric cancer (CGC)  and as a result of DRMR, 100 g of citrus fruit intake per day inhibits CGC by 40% .It is suggested that the intake of citrus fruit inhibits the development of CGC. This conclusion can be used as a primary prevention measure in the future when the incidence of CGC may be on the rise. Helicobacter pylori (Hp) infection [As one of the main types of primary cancers, gastric cancer is the fourth and fifth most common cancer worldwide in men and women, respectively, and has the highest incidence rate in Far East Asia, which includes Korea . The obsnfection -4. Howevnfection -7.In 2008, Bae et al. ,10.On the one hand, Fang et al.  publisheTherefore, this study aims to perform an adaptive meta-analysis on cohort studies published until December 2015 in order to overcome the shortcomings of the two previous SRs ,11 and iCohort studies that investigated the relationship between dietary intake of citrus fruit and gastric cancer risk were selected for this study. Related articles were selected through a three-step process involving searching in databases, examining titles and abstracts, and reviewing literature content, as suggested by preferred reporting items for SRs and meta-analyses .http://www.ncbi.nlm.nih.gov/pubmed) and Scopus (www.scopus.com). The final publication date was set to December 31, 2015. In addition, manual searching was done to check if a SR had already been published on the hypothesis in question, and more lists were created. Lists made manually and through the CDT were combined, and duplicated lists were eliminated.The data searching was done manually rather than electronically. This is because two SRs ,11 had bFor the screening stage, studies were excluded if they fell into the following criteria based on their titles and abstracts: (1) laboratory experiments, (2) expert/SRs, and (3) descriptive studies. Studies that passed the screening process then moved on to the third stage for eligibility assessment. Copies of each study were collected and evaluated, and cohort studies that met the following criteria were sequentially excluded: (4) analytical studies that did not provide information necessary for the meta-analysis, (5) case-control studies, (6) gastric cancer mortality-based cohort studies, and (7) cohort studies that were duplicated as a result of extending a follow-up period. The remaining studies after applying the seven exclusion criteria were selected as the final articles to be used in the meta-analysis.For each study, we looked at the cohort data source, the average length of follow-up period, the number of participants in the cohort as well as the number of cases who developed cancer, the method of investigating the quantity and unit of dietary citrus fruit intake, the adjusted relative risk (aRR) obtained after adjusting for confounders for each distribution of dietary intake and the corresponding 95% CIs, and whether any of the adjusted confounding factors controlled the total energy intake and history of Hp infection. In the cases where an aRR was reported separately based on sex  and anatomical region , we considered the aRRs as separate datasets instead of combining them into one.For aRR of each dataset to be used in the meta-analysis, the ‘interval collapsing’ method (ICM) was used instead of the ‘highest vs. lowest intake’ method (HLM). This is because ICM utilizes a greater amount of information than HLM and therefore improves statistical accuracy . In an Iwww.stata.com).In addition, in order to study the dose–response relationship for different quantities of citrus fruit intake, random effects dose– response meta-regression (RE-DRMR) was performed . For dosIn the subgroup analysis of each anatomical region, it was found that non-cardia gastric cancer had no statistical significance in relation to citrus fruit intake . However, citrus fruit intake had inhibitory effects on the development of cardia gastric cancer (CGC), and also had a statistically significant relationship with CGC risk  . Of the To summarize the results, dietary intake of citrus fruit reduced gastric cancer risk by 13%, and had an inhibitory effect on CGC in particular. Since the subgroup analysis and RE-DRMR analysis showed that citrus fruit intake has no statistical significance with respect to NGC, it can be inferred that citrus fruit intake inhibits the development of CGC. With antioxidant supplement found to have no anticancer effects -35, the The present study differs from the SR done by Fang et al.  in termsThis study has some limitations. First, none of the five selected cohort studies was adjusted for Hp infection. This may be due to the difficulty in performing Hp tests on all of the study participants during cohort establishment, with the importance of Hp infections being acknowledged only recently ,36. HoweTo summarize the results, dietary intake of citrus fruit inhibits the development of gastric cancer, especially CGC. With the growing trend in the incidence of CGC and the decreasing trend in the incidence of NGC , this fi


Meta-analysis of existing evidence provides a neutral association between allergies and CRC risk. Future well-designed prospective cohort studies should be conducted to better understand this association.A history of allergy or allergic condition has been reported to be associated with reduced risk of some types of malignancies. However, the understanding of this association for colorectal cancer (CRC) is controversial. We conducted a meta-analysis of CRC risk in individuals who had history of allergy compared to those without the history of allergic condition. Pumbed and Embase databases were searched for relevant studies. The adjusted relative risk (RR) and 95% confidence interval (CI) were pooled using the random-effects model. Nine studies, including 775, 178 individuals, were eligible for inclusion. The pooled estimate showed no significant association between history of allergy and CRC risk . Subgroup analyses confirmed the neutral association stratified by tumor location (colon:  Colorectal cancer (CRC) represents the third most common cancer and the fourth cause of cancer death worldwide, with an estimated 1.4 million new colorectal cancer cases and 693,900 deaths in 2012 globally , indicatAccumulating evidence suggests a pivotal role of allergic conditions in modulation of immune function , 13. We We performed this meta-analysis in accordance with the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) guidelines  Supplem.We retrieved 3,220 unique citations in the initial literature search and 39 potentially relevant studies for full-text review. After removing 30 studies, a total of 9 cohort studies met our inclusion criteria and were involved in the meta-analysis  Figure .The baseline characteristics of the identified studies are shown in Table Eight of the nine studies involved multicentric data , 25–28, n = 9 studies; adjusted RR 1.01, 95 % CI 0.88 to 1.17) . Significant reduction in heterogeneity was seen when studies involved in colorectum, hay fever, USA/Canada region, sample size less than 1000, questionnaire based assessment method and quality score > 6, indicating that the heterogeneity could be explained partly by these factors  suggested no publication bias. The results of the trim-and-fill method indicated that three studies might have been missing and the adjusted pooled RR was 0.93 (95 % CI 0.79 to 1.09) when inputing these three hypothised studies  and large sample size implied high statistical power. Second, the literature search, eligibility assessment, data extraction, and quality assessment were conducted independently by at least two investigators and one senior author. Third, in order to more conservatively calculate the risk estimate of CRC, we used a random effects model to combine data in overall population and subgroup analyses were thoroughly undertaken according to gender, tumor location, type of allergy and other baseline characteristics. Fourth, almost all of the studies selected were limited to those with population-based registries with representative samples, which provided better evidence than those provided by data derived from convenience samples. Finally, we assessed for publication bias using different approaches , giving sufficient evidence to confirm the evidence based on the current available studies.Several limitations should be addressed. First, as a meta-analysis of observational studies, there was a lack of the experimental random allocation of the intervention like randomized controlled trials which was an optimal method to test exposure – outcome hypotheses. Second, the adjusted variables varied among studies. They generally failed to adjust these following important risk factors for CRC in all studies: family history of CRC or other tumors, diet habits, smoking, alcohol use, nonsteroidal antiinflammatory drugs (NSAID) use and other related factors, which could be some of the influential factors of CRC. Moreover, three studies used SIRs to estimate CRC risk , 23, 28.The mechanisms of tumorigenesis associated with allery are controversial. There have been two contradictory theories proposed. One theory was that cancer risk could be reduced by some allergic conditions through immune surveillance, inducing immune reactions to remove malignant tumor cells, whereas the other theory proposes that allergic conditions can result in continuous tissue inflammation, damage and repair, which increases the risk of cancer , 34. A mIn summary, this systematic review and meta-analysis provide evidence that do not support a substantial a protective or harmful association between history of allergy and risk of CRC. Future well-designed prospective cohort studies should be conducted to better understand this association. Furthermore, studies should include analysis on more types of allergy in order to investigate if the effect of specific type of allergy on the tumorigenesis of CRC.A systematic search of literature was performed on January 26, 2016, in PubMed and Embase from the initial available date according to the Cochrane review guidelines. We used the following sets of Mesh/Emtree terms for searching: allergy, asthma, allergens and allergic rhinitis; colorectal neoplasms, colonic neoplasms and rectal neoplasms. The detailed search strategies are presented in Allergies or allergic conditions were defined as a self-report history of having been previously diagnosed by a physician that the respondent had asthma, hay fever, skin or food allergy, or any other allergy. All abstracts examining risk estimates of the association between history of allergy and CRC were screened for full-text review. Studies were considered eligible for inclusion if they satisfied the following criteria: prospective or retrospecitve human cohort studies; studies investigating the association between history of allergy and CRC and reporting the corresponding RRs or SIRs of CRC or sufficient data to calculate them.Two reviewers (J.Y. and A.T.) independently evaluated each eligible study and extract related data. Disagreements were resolved by discussion or by a third reviewer (H.Z.). A predesigned standardized data collection form regarding the baseline characteristics including the following items was used: first author, publication year, research country, inclusion period, study name, center involved, study design, sample size, mean/median age of included individuals, percent of male individual, type of allergy, exposure assessment method, measure of associations, outcome assessment, years of follow-up and adjusted variables.We used the Newcastle–Ottawa Scale (NOS)  to assesThe meta-analysis was performed abided by the preferred reporting items for systematic reviews and meta-analyses (PRISMA) statement . All staWe extracted the adjusted relative risks (RRs), odd ratios (ORs), hazard ratios (HRs), standardized incidence ratios (SIRs) and corresponding 95% confidence intervals (CIs), if available, from the included studies which were used to assess the association between history of allergy and CRC risk. For the low absolute risk of CRC, it is expected that the four measures of association can yield similar estimates of RR. Therefore, we can reasonably consider that the pooling of RR estimates can ensure the comprehensiveness of the meta-analysis and maximization of the statistical power , 39.2 statistic [2 > 50% was considered significant heterogeneity.Random-effects models were used to evaluate the pooled RR for the association between history of allergy and CRC incidence  because tatistic . An I2 >P value less than 0.05.Potential publication bias was evaluated through visual inspection of funnel plots combined with the Egger's regression test  served a