Monocyte chemoattractant protein-1 (MCP-1) plays an important role in the development of allergic inflammatory reactions by recruiting various immune cells, which is associated with many autoimmune diseases, but the association with the MCP-1-2518A/G gene polymorphism and lupus nephritis (LN) was still controversial in previous studies. Thus, we performed a meta-analysis to derive a more precise evaluation of the association between MCP-1 -2518A/G polymorphism and LN risk and evaluated influence of ethnicity and source of controls.A systematic review and meta-analysis that will be performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA). Relevant literatures dated to September 2016 were acquired from the PubMed, EMBASE, Cochran Library databases. A total of 961 LN cases and 1867 controls were extracted from 10 published case-control studies. We used odds ratios (OR) with 95% confidence intervals (CI) to assess the risk of LN with MCP-1-2518A/G.P < .01, OR = 1.42, 95% CI: 1.13–1.79 and A vs G P = .02, OR = 0.74, 95% CI: 0.58–0.95). Then the subgroup analysis showed MCP-1 -2518 A/G gene has a certain correlation with LN susceptibility in the American population , while no significant risk in Europeans and Asians.Our meta-analysis suggested that MCP-1-2518A/G polymorphism was associated with the risk of LN (GG vs AG+AA: The current meta-analysis suggests that the MCP-1-2518A/G polymorphism is associated with an increased risk of LN, especially in the American population. However, better-designed studies with larger sample sizes are needed to validate the results. With further investigation, the genetic variation of copy number variation and epigenetic changes were confirmed to be associated with the incidence of LN.Lupus nephritis (LN) is a result of systemic lupus erythematosus (SLE), as one of the most serious complications of SLE, which leads to the deposition of glomerular immune complex and the inflammation of renal glomerular injury. In addition, Singh et al found the level of urinary MCP-1 in LN patients was significantly higher than those without LN, and correlated to the severity of LN, which indicated MCP-1 might be a biological indicator of LN. Subsequently, Mohammed's research further confirmed the association of MCP-1 and susceptibility LN. However, Alharazy et al noted that there was no certain association between the level of MCP-1 and the LN. Furthermore, the gene phenotype of MCP-1 -2518A/G was not associated with the incidence of LN. Obviously, the association with the MCP-1-2518A/G gene polymorphism and LN was still controversial. Considering the relatively small sample size in most studies, it is possible to perform a quantitative synthesis of the evidence with rigorous methods. Here, we performed a meta-analysis on 10 published case-controls to derive a more precise evaluation of the association between MCP-1-2518A/G polymorphism and LN risk.Monocyte chemoattractant protein-1 (MCP-1) plays an important role in the development of allergic inflammatory reactions by recruiting various immune cells, which is associated with diverse autoimmune diseases.22.1The keywords “monocyte chemoattractant protein-1,” “MCP-1,” “lupus nephritis,” “systemic lupus erythematosus,” and “polymorphism” were searched in 3 databases  for entries until September 2016. References of the retrieved publications were also screened. All eligible studies were retrieved, and their bibliographies were checked for other relevant publications. Only published studies with full-text articles were included. When overlapping articles were found, we only included the publications that reported the most extensive information.2.2Literatures fulfilled the following criteria: published in English; experimental subjects were the LN patients, and the control groups were healthy people; case-control studies of LN with MCP-1-2518 A/G polymorphism; sufficient published data for estimating an odds ratio (OR) with 95% confidence interval (CI); case control groups genotype conformed to the Hardy Weinberg (H-W) balance.2.3Two investigators independently reviewed the articles to exclude irrelevant and overlapping studies. The results were compared, and disagreements were resolved by discussion and consensus. We extracted the following information from each study: first author's surname, publish year, ethnicity, and the number of cases and controls for each genotype, gene detection method, and control source.2.4χ2-based Q statistic test was performed to evaluate the between-study heterogeneity of studies. If the heterogeneity was not significant , then the fixed-effect model can be performed, otherwise, the random-effect model. Subgroup analyses were conducted among variables, such as ethnicities and source of controls. Sensitivity analysis was conducted by removing 1 data set at a time to identify individual study's effect on pooled results and test the reliability of results. Funnel plots were used to access the potential publication bias by the method of Egger linear regression test. All analyses were performed by Stata  and Review Manager , using 2 side P values.The odds ratio (OR) with 95% confidence interval (CI) was used to assess the strength of association between MCP-1-2518 A/G polymorphisms and LN risk in 5 genetic models . The 33.1–17 were adopted in the end, including 961 LN cases and 1867 controls met the including criteria. The populations were from the Americas, Europe, and Asia. A classic polymerase chain reaction–restriction fragment length polymorphism assay was performed in all studies. The characteristics of these studies and the quality scores were shown in Table Our search strategy identified 34 potentially relevant studies, and a total of 10 literatures3.2P = .00, OR = 1.42, 95% CI: 1.13–1.79 and A vs G: P = .02, OR = 0.74, 95% CI: 0.58–0.95, respectively). Subsequently, subgroup analysis was performed based on the ethnicity , which suggested the MCP-1-2518 A/G gene has a certain correlation with LN susceptibility in the American population GG vs AA: P < .01, OR = 5.70, 95% CI: 2.09–15.50, GG vs AG+AA: P < .01, OR = 3.31, 95% CI: 1.97–5.54, GG+AG vs AA: P < .01, OR = 2.86, 95% CI: 1.14–7.18, and A vs G: P < .01, OR = 0.43, 95% CI: 0.24–0.79; Figs. The evaluation of association between MCP-1-2518A/G polymorphism and LN risk was presented in Table 3.3We used sensitivity analysis to estimate individual study's influence on the pooled OR, and the result of sensitivity analysis showed no other single study influenced the summary OR qualitatively, suggesting stability of the meta-analyses Fig. .3.4t = −0.66, P = .53).Funnel plots are shown in Fig. 4The present meta-analysis, including 961 LN cases and 1867 controls from 10 published case-control studies, showed that the MCP-1-2518 A/G was associated with LN risk. When stratified by different ethnicities, we found no significant association with European and Asian populations, which mainly occurred in the American population. This difference may be due to the genetic background of various ethnic populations and the environmental exposure. Further research on MCP-1 gene can better clarify the biological characteristics of LN, which can help us to detect, prevent, and treat LN. The results from the present meta-analysis have a certain differences. The differences may be due to the publication of the latest case-control studies, but it is still more likely to be further confirmed based on larger population and more ethnicities studies. In many comparisons, there was heterogeneity, which brought the potential interference to the study. Although we minimized the likelihood by performing a careful search for published studies, using strict criteria for study inclusion, precise data extraction, and careful data analysis, significant between study heterogeneity existed in some comparisons. Then the subgroup analysis to a certain extent reduced the heterogeneity, but it was not entirely able to completely control the heterogeneity at a very low level. In the method of statistical analysis, we used the random-effect model to deal with the obvious heterogeneity . Since the included studies were multicenters and regions, the heterogeneity of sources may come from the genetic, geographical environment, and living habits, etc.Previous meta-analysis suggested that the polymorphism of MCP-1 gene was not associated with the incidence of LN, but which was associated with the susceptibility of LN in Caucasian population. found AA genotype might be a biomarker for the patients with SLE developing into LN, and the AA genotype was associated with the onset of SLE in Caucasians. Furthermore, a trend toward an association between A allele and LN risk was observed in Caucasians. The results of gene polymorphisms suggested the susceptibility of certain populations to LN, and the direct application is screening for this population. In addition, a number of articles reported the relationship between polymorphisms and LN treatment, including efficacy and side effects,,20 and drug interactions. However, few studies have reported that MCP-1-2518 A/G polymorphism is involved in the therapeutic implications of LN.The reasons for the difference found between the Caucasians and Asians population may be multifaceted. The difference may be related to the presence of other environmental factors and the distribution of the gene. LN is a result of SLE. A previous researchThe pathogenesis of LN has many aspects, and is regulated by many genes and signals. The analysis of a single nucleotide variation may not be able to fully elucidate the genetic association of LN, which may involve multiple genes. In addition, this study has the following limitations: First, the number of sample sizes included in the study was still not enough, it may cause some unstable results; Second, only English-language studies that were included in this meta-analysis might have led to publication bias, and the exclusion of unpublished data was generally associated with an overestimation of the true effect; Third, controls were not uniformly defined, while our result was based on unadjusted estimates.In conclusion, our results suggested the MCP-1 gene-2518 A/G polymorphism was correlated with the presence of LN. Subgroup analysis found the genetic susceptibility was mainly associated with the American population, whereas there was no significant association between European and Asian populations. In future studies, more and larger case-control studies are needed to obtain further confirmation. Larger samples among different ethnicities, especially more sophisticated gene–gene and gene–environment interactions, should be considered in future studies, which should lead to better, comprehensive understanding of the association between MCP-1-2518 A/G polymorphism and LN risk.


Stratified analysis based on ethnicity found that LIPC rs493258 polymorphism had a significant association with the decreased risk of the disease in the Caucasian population, but not in the Asian population. For late AMD, significant associations of the rs493258 polymorphism with a lower risk of this disease were also observed in the allelic genetic model . This meta-analysis demonstrates that the T allele in the LIPC rs493258 polymorphism was significantly associated with the risk of any and late AMD. The associations of the locus with early and late AMD risk in various populations need further exploration.The purpose of this study was to evaluate the association of the hepatic lipase (LIPC) rs493258 polymorphism and susceptibility to age-related macular degeneration (AMD). A systematic search in PubMed, EMBASE, and ISI web of science databases was performed to identify eligible published studies without language restrictions up to April 2016. Pooled odds ratios (ORs) with 95% confidence intervals (CIs) in different stages of AMD were estimated under different genetic models using meta-analytic methods. Seven studies comprising 20,559 cases and 17,200 controls met the inclusion criteria and were included in the meta-analysis. The LIPC rs493258 polymorphism showed a significant association with a lower risk of AMD under the allelic model . Significant relationships between the variant and AMD were also observed in other genetic models (OR ranging from 0.71 to 0.86, all  With a high incidence rate of 8.7% across the globe, age-related macular degeneration (AMD) is identified as the leading cause of irreversible visual loss among elder individuals in developed countries . It is eAMD is a progressive disease caused by a multitude of factors, including environmental and genetic factors. Although many epidemiological studies have linked age, sex, cigarette smoking, alcohol consumption, light exposure, and food habits to the risk of AMD, the pathogenesis of AMD still remains vague ,5,6. MorTherefore, we conducted a meta-analysis to evaluate the association between LIPC rs493258 polymorphism and AMD risk. Moreover, stratified analyses were also performed to explore the association at different AMD stages and in different races.A comprehensive literature search of PubMed, EMBASE, and the ISI Web of Science databases was conducted to identify the relevant studies. The search strategy was based on a combination of the terms  and . The searches were limited to human studies with no language restrictions. In addition, we also checked all of the references of relevant reviews and retrieved eligible articles. When data were inadequate for necessary analysis, corresponding authors were contacted, in most cases successfully, for the retrieval of missing data.The included studies fulfilled the following inclusion criteria: (1) the original report evaluated the association of the LIPC rs493258 polymorphism with the risk of early or late AMD; (2) the study design was limited to cohort, case-control, or cross-sectional design; and (3) the study presented available data which provided an odds ratio (OR) with a 95% confidence interval (CI) under different genetic models or sufficient data to estimate them.The following information was independently extracted from each study included in the present meta-analysis by two investigators (Yafeng Wang and Mingxu Wang): the first author’s last name, year of publication, study design, sample size, the ethnicity of the study population, the mean age, proportion of male, definition of cases and controls, classification criteria, genotype distributions in cases and controls, genotyping methods, and minor allele frequency and Hardy-Weinberg equilibrium (HWE) among controls. The results were compared, and any inconsistencies were resolved through discussion.The quality of the included studies was independently appraised with the nine-item Newcastle Ottawa Quality Scale (NOS), a widely used tool for the quality assessment of observational studies . Three bI2 index were employed to assess the between-study heterogeneity. A p value < 0.10 for the Q test showed an existence of heterogeneity, and an I2 value >50% indicated a statistical significance. When heterogeneity was negligible, the fixed-effects model was chosen to evaluate the pooled estimate of risk; otherwise, the random-effects model was performed. Subgroup analysis and meta-regression were performed to explore the potential sources of heterogeneity between studies. In addition, sensitivity analysis by sequentially removing one study at a time was performed to confirm the stability of the results. Publication bias was assessed by the Begg test and the Egger test [p value < 0.05 was considered significant.HWE was tested by the chi-square test. The strength of associations between the LIPC rs493258 polymorphism and the risk of AMD was assessed by odds ratios (ORs) with 95% confidence intervals (CIs). We explored the association of the LIPC rs493258 polymorphism with AMD risk under the allelic (T vs. C), dominant ((TT + TC) vs. CC), heterozygous (TC vs. CC), and homozygous (TT vs. CC) genetic models, respectively . The Cochran’s Q statistic and ger test ,13. All A total of 212 articles were identified by the primary literature search . After tp > 0.05).Characteristics of the included studies are summarized in I2 = 0%, p = 0.89). The significant relationships between this variant and the AMD risk were also observed in other genetic models  vs. CC: OR = 0.80, 95% CI = 0.74–0.86, p = 0.24) and Egger’s regression test (p = 0.17).Sixteen studies evaluated the associations between the LIPC rs493258 polymorphism and risk of AMD in the allelic (T vs. C) genetic model, and 14 studies in the other three genetic models. When these studies were pooled in the present meta-analysis, the results showed that the rs493258 variant showed a significant summary OR under an allelic model of T vs. C .We subsequently evaluated the association of the rs493258 variant with late AMD in six studies. Four studies assessed the associations of the rs493258 variant with late AMD in the allelic genetic model, and two studies were in the dominant, heterozygous, and homozygous genetic models. The results found that the T allele in the rs493258 variant significantly decreased the risk of late AMD  and genetic factors. Thus, further research studies should be conducted if individual data are available, which could adjust other covariates including age, sex, environmental factors, and genetic factors. Third, only published studies were included in our meta-analysis; thus, the potential publication bias was also a concern. Although the apparent publication bias by statistical tests was not observed, we could not completely rule out this problem because a much greater number of studies is necessary in order to detect it adequately.The meta-analysis demonstrated the association of LIPC rs493258 polymorphism with AMD. Carriage of the LIPC rs493258 T allele can decrease the risk of developing AMD. Moreover, this polymorphism can decrease genetic susceptibility for AMD in Caucasians. However, only one study has assessed the association of LIPC rs493258 polymorphism with early AMD, and the number of included studies based on Asians was relatively limited; therefore, further well-designed large-perspective cohort studies with interethnic populations are required to evaluate these associations at different stages of AMD.


The purpose of this study is to summarize the currently available evidence regarding the concerned issue by performing a comprehensive meta-analysis. Relevant publications reporting the association of metformin use with survival of lung cancer patients with diabetes were electronically searched to identify eligible studies. The meta-analysis was performed with hazard ratios (HRs) and 95% confidence intervals (95% CIs) as effect measures for disease-free survival(DFS) and overall survival(OS) estimates. A total of 17 individual studies from 10 publications were included in the meta-analysis. Overall, the results revealed a significant association of metformin use with a better survival of lung cancer patients with diabetes. The subgroup analyses showed similar association in Asian region but not in Western region. Such association was also presented in small cell lung cancer  and in non-small cell lung cancer. Analyses stratified by treatment strategy showed a reduction in the risk of cancer-related mortality in patients receiving chemotherapy but not in patients receiving chemoradiotherapy. The meta-analysis demonstrated that metformin use was significantly associated with a favorable survival outcome of lung cancer patients with diabetes. According to the epidemiologic and clinical studies, approximately 8% to 18% of individuals survived with a diagnosis of cancer accompanying with diabetes, probably due to the shared risk factors of the diseases, such as greater body mass index, smoking, and the lack of exercise and their growing global prevalence . The conMetformin as one of the most commonly prescribed oral antidiabetic agents for type 2 diabetes may acts by inhibiting hepatic gluconeogenesis , reducinUp till now, a number of epidemiological studies have investigated the prognostic significance of metformin use in lung cancer patients with diabetes, but obtained inconsistent results. Tan et al. reported that metformin may improve chemotherapy outcomes and survival for advanced non-small cell lung cancer (NSCLC) patients with type 2 diabetes . Xu et aThe participant flow diagram for the included studies is shown in Figure p = 0.513), which suggested no evidence of publication bias. Stratification analyses based on region revealed a reduced risk of recurrence or progression in lung cancer patients with diabetes for metformin use in Asian region  with a fixed effect model. However, no such significant association was observed in Western region. In the subgroup analysis by treatment strategy, the results showed metformin administration was significantly associated with decreased risk of recurrence or progression in patients receiving chemotherapy , while surprisedly, no such significant survival benefit was reveled in patients receiving chemoradiotherapy. Furthermore, we also performed subgroup analyses by subtype of lung cancer. Our results demonstrated a significant association between metformin administration and decreased risk of recurrence or progression .The pooled HR for association between DFS of lung cancer and exposure to metformin is shown in Table p = 0.763). Analyses stratified by treatment strategy showed a reduction in the risk of cancer-related mortality in patients receiving chemotherapy, while no reduced risk was presented in patients receiving chemoradiotherapy. The stratification analyses by region showed that metformin was associated with reduced death risk in Asian region  but not in Western region. In the subgroup analyses based on subtype of lung cancer, the relative survival benefit associated with metformin remained in both SCLC and NSCLC subtype .Considering the effect of metformin administration on the OS, as shown in Table Considering the large variations across studies, we performed sensitivity analysis by sequential omission of individual studies to recalculate the pooled HR for the remainder of the studies, and none of the exclusions of a specific study would substantially alter the trend of our primary pooled results, showing a significant association of metformin administration and improved DFS. Also, the pooled result was not materially affected in the sensitivity analysis concerning the OS.Our meta-analysis pooling currently available observational studies was designed to comprehensively and quantitatively evaluate the relationship between metformin use and the clinical outcome of lung cancer patients with diabetes, demonstrating a significant association of metformin therapy with improved DFS and OS.Increasing evidence has strengthened the effects of metformin on prevention and treatment of lung cancer in recent years though the exact molecular mechanism for the anti-cancer role of metformin has not been fully unveiled to date. In vitro studies found that metformin inhibited the proliferation and cell cycle, and induced apoptosis in different lung cancer cells . The nudIn the present meta-analysis, metformin treatment was demonstrated to prolong the DFS and OS of lung cancer patients with diabetes, which was consistent with the previous meta-analysis . In straIn spite of the advantages, caution is called for the appropriate interpretation of our findings with the consideration of several potential limitations existing in the study. First, although adjusted estimates were largely utilized in the meta-analysis, the adjusted confounders varied across studies and some potential residual or unknown confounders might not be fully adjusted, which may affect the survival of cancer patients and thus bias the effect estimate of metformin on survival. We also used unadjusted estimates for the study to minimize potential bias because we failed to obtain the individualized primary data from the included studies, though we have tried to contact the authors for data request. Second, details including metformin dose, and other significant information relevant to cancer prognosis, especially including tobacco use , beside In summary, our meta-analysis demonstrated a significant association of metformin use with a favorable survival outcome, suggesting a potentially available choice of treatments for lung cancer patients with diabetes. Meanwhile, the current study reveals the need for more blind randomized controlled clinical trials to further assess the efficacy of an antidiabetic regimen, metformin in the treatment of lung cancer.We searched PubMed, Embase, Web of Science, Cochrane Library, Wan Fang, China National Knowledge Infrastructure, and the Chinese BioMedical Literature databases for relevant studies investigating the association between metformin use and the survival outcome of lung cancer patients with diabetes up to March 2016 without language restrictions. We developed a search strategy using the following terms in various combinations: [“metformin” OR “biguanides”] and [“lung cancer” OR “lung carcinoma”]. We also screened the reference lists of identified studies and reviews for additional relevant studies.Studies were considered eligible if they fulfilled the following criteria: (1)case-control study or cohort study assessing the association of metformin use and prognosis of lung cancer patients with diabetes. (2) studies contained time to event data  pertaining to association between metformin use and survival of lung cancer patients with diabetes. Studies with a small sample size and low study quality or no time to event data provided were excluded. In case of duplication, only the study with the most complete information and largest sample size was selected.For each included study, the following information was independently extracted by two investigators: first author, year of publication, region, country, study design, reference of group, subtype of lung cancer, clinical stage, sample size, treatment choice of patients, adjusting variables, HRs with their 95 % CIs for DFS and OS. For observational studies, the results from both unadjusted and adjusted models  were likely selected. If several estimates were reported in the same article, we select the most fully adjusted estimate . If an article reported multiple estimates by subgroup only, these estimates were entered separately into our relevant meta-analysis data set. Any disagreement was resolved by discussion or consultation with a third reviewer to reach a consensus. The Newcastle-Ottawa quality assessment scale was applied in the quality assessment of included studies, concerning some of important factors described as the previous studies , 38.http://www.cochrane.org/software/revman.htm). The significance of the pooled HR was determined by Z test and p < 0.05 was considered significant. Inter-study heterogeneity across studies was assessed using the Cochran's χ2-based Q test and the I-squared test [2 < 50% indicated the absence of heterogeneity, then the fixed effect model was selected to calculate the pooled HRs [p < 0.05 indicated the presence of publication bias) [Pooled HRs and associated 95%CIs were calculated using a fixed or random effect model according to the heterogeneity across the studies with Review Manager version 5.2 software  , 43, usi


CI = 0.78 – 0.94), 0.78 (95% CI = 0.66 – 0.86), 4.00 (95% CI = 2.47 – 6.49), 0.16 (95% CI = 0.08 – 0.30), 25.86 (95% CI = 9.29 – 71.95), and 0.90 (95% CI = 0.87 – 0.92), respectively, indicating that miR-205 is a useful biomarker for diagnostic of lung cancer. The subgroup analysis further demonstrated that miR-205 had an excellent overall accuracy for detection with tissue samples compare with blood samples. For the prognostic meta-analysis, the pooled outcome of the disease-free survival and recurrence-free survival analyses revealed that increased miR-205 levels had a protective role in the prognosis of patients with lung cancer . In conclusion, miR-205 may be a promising biomarker for detection, predicting the recurrence of patients with lung cancer.MicroRNA-205 (miR-205) was revealed as a novel diagnostic and prognostic biomarker for lung cancer, but the results in the published papers were inconsistent. This meta-analysis aimed to investigate the diagnostic and prognostic roles of miR-205 in patients with lung cancer. Totally, 16 eligible articles were included, among which 10 articles investigated the diagnostic value of miR-205, 5 articles examined its prognostic values, and 1 article studied both diagnostic and prognostic values. For the diagnostic meta-analysis, the pooled sensitivity, specificity, positive likelihood ratio, negative likelihood ratio, diagnostic odds ratio, and the overall area under the curve of miR-205 for patients with lung cancer were 0.88 (95%  Lung cancer is the leading cause of cancer-related death among men and women worldwide with only a low 15% overall 5-year survival rate and a high recurrence rate . Early dMicroRNAs (miRNAs), as endogenous and non-coding small RNAs, suppress gene expression by binding to 3′- untranslated region (UTR) of targeted messenger RNAs (mRNAs), leading to gene degradation or translation suppression , 11. DysMiR-205, located in a lung cancer-associated genomic amplification region at 1q32.2 , 19, parA total of 662 potentially relevant articles were obtained from electronic databases and other sources, and 499 of them were remained after removing 163 duplications. Of the remained articles, 460 articles, such as letters, reviews and meta-analysis, or unrelated to the research topic, were excluded after reviewing titles and abstracts. Of the remaining 39 full-text candidates, 23 potential articles were further excluded, which did not have sufficient data, or were unrelated to cancer diagnosis or prognosis. Finally, 16 eligible articles , 23–37 wTotally, 564 lung cancer patients (mainly composed of SCC patients) and 667 controls  were included in this meta-analysis. Because Hamamoto et al  investigI2 values of pooled sensitivity and specificity were 89.45 % and 89.80 %, respectively, indicating that there is a significant heterogeneity among the included studies.The pooled results of sensitivity, specificity, diagnostic odds ratio (DOR), and the area under the summary receiver operator characteristic curve (AUC) were 0.88, 0.78, 25.86, and 0.90, respectively Table  and 4, rNomogram of Fagan was used to estimate the clinical diagnostic values of miR-205 in detection of lung cancer. In details, for any people with a pre-test probability of 25 % of patients with lung cancer, positive results of miR-205 show the post-test probability of correctly diagnosing cancer would rise to 57%, while negative results of miR-205 mean the post-test probability would drop to 5 % , which indicated that the heterogeneity was not caused by the threshold effect.The threshold effect caused by differences in the sensitivity and specificity is a major source of heterogeneity in diagnostic tests. Hence, spearman correlation coefficient of sensitivity and specificity was calculated to assess threshold effect. According to the analysis, spearman correlation coefficient in total 12 studies was 0.33, with a Subgroup analysis was performed based on ethnicity (Asian vs. Non-Asian), specimens (tissue vs. blood), and sample size (≤ 100 vs > 100) Table . There wSensitivity analysis was conducted and 1 outliner was found Figure . After eP value of Deek's test was 0.54.Deeks’ funnel plot was performed to evaluate the publication bias of the included studies. The funnel plots of diagnostic results indicated no obvious publication bias in this diagnostic meta-analysis Figure . The P vThe main features of the 6 included studies for prognosis are listed in Table P = 0.075, I2= 53 %). Therefore, the pooled hazard ratio (HR) was summarized by using a random-effect model. Our results failed to demonstrate any significant association between miR-205 expression and OS . Therefore, the pooled HR was summarized by using a fixed-effect model. Our results indicated that the increased miR-205 expression had a protective role in the prognosis of patients with lung cancer . Our finding was supported by Zhang et al [To further solve the controversial results of miR-205 in prognosis of lung cancer, for the first time, we conducted this meta-analysis to evaluate the relationship between miR-205 expression and the prognosis of patients with lung cancer. Although several studies supported that increased miR-205 expression may predict a worse OS , 49, ourThis study had several strengths. To our knowledge, we performed a meta-analysis to evaluate the diagnostic and prognostic efficiency of miR-205 for patients with lung cancer for the first time. Moreover, different sample sources were investigated, with the aim of identifying the most suitable one for clinical application. In addition, we used comprehensive methods, including strict literature screening and quality evaluation process, to reduce bias.Limitations do exist in the meta-analysis. Firstly, the blind design was not used in some diagnostic studies, which affects the diagnostic accuracy. Secondly, the sample size in this meta-analysis was relatively small. Thirdly, considering heterogeneity existed among these studies.In summary, this study demonstrated that miR-205 may be used as a promising biomarker for diagnosis and prognosis for lung cancer. Our findings require further evaluation in future large-scale prospective studies.A systematical computerized search was performed for relevant publications that investigated the roles of miR-205 in the diagnosis and prognosis of lung cancer in Embase, PubMed, Web of Science databases, and Ovid platform  for the terms  and (“microRNA-205” or “miRNA-205” or “miR-205” or “hsa-miR-205”). We further manually searched the bibliographies of articles to identify the missed suitable articles.The included studies in this meta-analysis satisfied all of the following inclusion criteria: (1) patients with lung cancer; (2) the expression levels of miR-205 in blood, serum, plasma, cytologic samples, or lung cancer tissues; and (3) sufficient data on the association between miR-205 expression levels and lung cancer diagnosis or prognosis. The exclusion criteria were as follows: (1) duplicate publications; (2) review paper, case report, letter, and meta-analysis; (3) unqualified data; and (4) non-English publications.P value were directly extracted from the studies, or extracted from the Kaplan-Meier curves by using a method as Tierney's study [Two reviewers (Jing-Hua Li and Shan-Shan Sun) independently carefully reviewed the full text and extracted the relevant data. A third person (Ning Li) resolved the differences until all arrived at a set of similar statements. After that, the following data were extracted from each study: name of the first author, published year, country, study population characteristics , and relevant data for meta-analysis. For diagnostic studies, data of two-by-two tables were extracted, including false negatives (FN), true negatives (TN), true positives (TP), and false positives (FP) etc. For prognostic studies, HR of miR-205 for OS, RFS and DFS, with corresponding 95% CIs and 's study .The methodological quality assessment of diagnostic study was conducted by independent team members according to the guidelines of the QUADAS-2 tool , which iP value (< 0.05) for heterogeneity or high I-squared (> 50%) suggests presence of heterogeneity caused by non-threshold effect. If the heterogeneity caused by non-threshold effects existed, stratified analysis  would be used to explore the sources of heterogeneity. The Fagan's nomogram was conducted to explore the clinical diagnostic value of miR-205 in detection of cancer. The Deek's funnel plot method was used to explore publication bias, with P < 0.05 indicates obvious publication bias.Meta-analysis was conducted using STATA 13.0  and Review Manager 5.3  software. The bivariate meta-analysis model was employed to calculate the pooled results of sensitivity, specificity, PLR, NLR, and DOR along with their 95% CIs, and generate the summary receiver operator characteristic (SROC) curve. The AUC represents an analytical summary of test performance , 54. MorP value (< 0.05) for heterogeneity or high I-squared (> 50%) suggests that the random-effects model would be applied, otherwise, the fixed-effects model was used. Moreover, the Begg's test and Egger's test were used to check the publication bias. All values of the P < 0.05 were considered represent statistical significance.For the prognostic meta-analysis, HRs and their 95% CIs were used to assess the impact of miR-205 expression on survival of patients with lung cancer. Cochran’s-Q and I-squared statistics index were used to assess the heterogeneity of the pooled results. A low


Antidiabetic medication may modify the incidence of hepatocellular carcinoma (HCC). We aimed to compare the use of different antidiabetic strategies and the incidence of HCC. PubMed, Embase.com and Cochrane Library databases were searched up to 31 October 2015 and randomized controlled trials (RCTs), cohort studies or case-control studies were included for our analyses. A total of thirteen studies enrolling 481358 participants with 240678 HCC cases who received at least two different strategies were retrieved in this analysis. Direct comparisons showed that use of metformin  was associated with a significant risk reduction of HCC, while insulin  may significantly increase the risk. Indirect evidence also suggested that insulin  was associated with a significantly increased risk of HCC. Additionally, metformin was effective in reducing the risk of HCC when compared with sulphonylurea  and insulin . Notably, metformin was hierarchically the best when compared with other antidiabetic therapies for the prevention of HCC. In summary, available evidence suggests that metformin was the most effective strategy to reduce HCC risk when compared with other antidiabetic interventions. The incidence and mortality rate of hepatocellular carcinoma (HCC) has significantly increased over recent decades. With a poor 5-year survival rate, HCC has become the third most common cause of cancer death worldwideDiabetes mellitus (DM) has been proposed as a potential risk factor for HCC5678in vitro and in vivo experimental studies has shown that insulin, it’s analogs and oral insulin secretagogues (such as sulfonylureas), which may contribute to hyperinsulinemia, are suggested to increase the incidence and progression of cancer1415On this basis, recent research has suggested that the use of antidiabetic medication may modify the risk for developing HCC in several different ways11Currently, there is still no high-quality RCT assessing the potential impact of different classes of antidiabetic pharmacotherapies on modifying the risk of developing HCC in DM patients. Due to the discrepant results in the literature and limited epidemiological evidence for the precise relationship between DM treatment and the risk of HCC, this study sought to systematically review the literature to evaluate, quantify, and summarize the association between the use of different anti-diabetic medications and the development of HCC. To obtain a better understanding of this issue, we performed a network meta-analysis of available clinical studies to investigate the association between antidiabetic medications and the risk of developing HCC.The participant flow diagram for study inclusion in the meta-analysis is shown in 151617181920212223242526Five different comparisons including observation were performed in pairwise meta-analysis, with a lack of direct comparison between TZDs and observation alone. The weighted RRs for the occurrence of liver cancer were calculated for each comparison, the geometric distribution of which is displayed in 2 statistic. Overall, statistical heterogeneity in our analysis was moderate, although for some comparisons 95% confidence intervals (CIs) were wide and included values indicating very high or no heterogeneity. As illustrated in 2 values higher than 75% were recorded for 3 comparisons, that is, metformin versus insulin, metformin versus sulphonylurea, TZDs versus sulphonylurea, with remaining comparisons lower than 40%. In addition, no publication bias was found for Begg’s rank correlation test and Egger’s test among these comparisons of different antidiabetic therapies.With respect to statistical heterogeneity, it was estimated in two of the comparisons by the IWe summarize the results of the random-effects network meta-analysis for HCC rates in The probabilities of best treatment for each strategy were ranked at each of the five possible parameters . In agreAccording to the results of traditional pairwise  and netwThis analysis, based on one RCT and 12 observational studies, showed that metformin had a chemopreventive effect on the incidence of HCC when compared with observation alone, while insulin was associated with a statistically significant increase in HCC risk. In addition, the probabilities of best treatment for each strategy suggested that metformin was the best, TZDs were the second best, sulphonylurea was the third best, and insulin was ranked the lowest in the prevention of HCC. Moreover, there was no inconsistency or publication bias in our network meta-analysis.According to our current knowledge and available data, glucose lowering drugs can variably modify the incidence of HCC. Our finding is consistent with the current understanding that exogenous insulin therapy or insulin secretagogues may be associated with an increased incidence of hepatoma and a higher mortality because of cirrhosis and HCC1321143033Notably, the association of DM and cancer is implicated in insulin resistance and hyperinsulineamia37vitro and in vivo studies supports the concept that PPAR-γ deficiency produce an environment prone to tumorigenesis, while PPAR-γ overexpression may inhibit HCC growth. Thus, PPAR-γ could exert beneficial effects against HCC and may represent as an anti-tumorigenic and therapeutic targetConversely, metformin, together with other insulin sensitizers, may counteract insulin resistance and consequent hyperinsulinemia, and appears to be associated with a lower cancer risk4042Some methodologic issues should be acknowledged which could be regarded as limitations in our study. First, our findings are primarily based on data derived from observational studies, which are inevitably prone to bias and confounders. Given the retrospective or hospital-based nature of some included studies that could lead to an overestimate of effect, further prospective studies with high quality analyses are needed to validate the effect of antidiabetics on such outcomes. Secondly, details on dosage, duration of antidiabetic therapy as well as full information on other residual confounders were incomplete. Thus, the findings provided by this meta-analysis should be viewed with some caution. Thirdly, statistical heterogeneity was moderate in the meta-analyses of direct comparisons, while no substantial inconsistency was found in the network meta-analysis. The diversity in study populations, comparators, and study design is responsible for a substantial heterogeneity in effect estimates across studies. Lastly, as the number of included studies is small, a bias (including bias of indication) should be considered in interpreting the results described.Besides these limitations, our meta-analysis has several strengths. Our study is the first Bayesian network meta-analysis performed on the potential tumor-modifying effect of antidiabetics. Based on the available literature and current knowledge, we have constructed a comprehensive and complete picture of the relationship between different antidiabetic medications and the incidence of HCC. In the absence of available studies directly comparing all eligible treatments, we have to rely on indirect comparisons of multiple treatments (such as the comparison arm of observation vs. TZDs in these studies). For instance, an indirect comparison of A with B can be made by integrating the information derived from studies of A vs. C and studies of B vs. C. Thus, the method of network meta-analysis may increase statistical power by incorporating evidence from both direct and indirect comparisons across all interventions. To ensure the accuracy and reliability of our results, this study compared the pool effect estimates of all available antidiabetic therapies, even when no head-to-head studies existed. For the included studies, major confounding factors have been controlled which include age, sex, race, socioeconomic status, body mass index, smoking, ethanol intake, hepatitis C virus infection, hepatitis B virus infection, cirrhosis, and DM duration. Data quality was reliable and the total statistical heterogeneity was moderate, with no small study effects or publication bias. Lastly, a rankogram of insulin, sulfonylureas, metformin and TZDs for chemoprevention of HCC provides a formal rank order for suggested treatment strategies for diabetic patients.In summary, our network meta-analysis provides a complete picture of the HCC-modifying effects of different antidiabetic medications by using Bayesian analytical approach. Specifically, metformin and TZDs have beneficial effects on HCC incidence, while insulin or sulphonylureas therapy may be associated with a higher risk of HCC. In light of potential confounding factors in observational studies, there remains a need for more well-designed randomized controlled trials, together with pathophysiological studies, to elucidate the potential role and the clinical efficacy of metformin and TZDs anticancer agents and to describe the details of their biological mechanism of action. Our analysis may contribute clinical decision making regarding appropriate antidiabetic treatment for patients with DM with a high risk of HCC.The protocol for this systematic review was performed in agreement with the Preferred Reporting Items of Systematic Reviews and Meta-Analyses (PRISMA) statementCriteria for inclusion of an article in this meta-analysis were as follows: (1) RCTs and epidemiologic studies, including cohort study or case-control study; (2) the objective was designed to evaluate the relationship between hypoglycemic agents and HCC incidence; (3) patients with no prior diagnosis of primary liver cancer or other cancers most likely to metastasize to the liver; (4) risk estimates should be provided or could be estimated with sufficient information. Duplicated or overlap reports were eliminated according to the same title, author list or publication date to avoid giving double weight to estimates derived from the same research or population. We then reviewed the full articles passing the first screening of titles/abstracts.Two investigators  independently reviewed the full manuscripts of eligible studies and extracted information into an electronic database: the first author, year of publication, country or area, patients’ characteristics, study design, type of hypoglycemic agents, study time or follow-up time, adjusted confounders, number of HCC cases and controls. Any discrepancies regarding the extraction of data were resolved by consensus and arbitration by an additional investigator (Zheng MH). When relevant information was unclear, or when some needed data was unavailable directly from the study, the original authors were sought for eligible data by email.To assess the risk of bias in the observational studies, we used the Newcastle–Ottawa Quality Assessment Scale with some modifications to match the needs of this study, which included the following items: patient selection, comparability of antidiabetic medication and observation group, and assessment of outcome. In terms of RCT, we used the Cochrane risk of bias tool to determine the quality. The quality of the methodology was independently assessed by two investigators .2 statistics. A formal confirmation of heterogeneity was then obtained by referring to the I2 statistic, judging values of less than 25% to be minimal, 25% to 49% to be moderate, and 50% or greater to be substantial. Nevertheless, the node splitting method was also utilized to estimate inconsistency of the model, which separated evidence into direct and indirect and compared them respectively. Small study effects were explored by inspecting comparison-adjusted funnel plots. Publication bias was assessed by Begg’s rank correlation test and Egger’s test among comparisons of different antidiabetic therapies. Egger’s test and Begg’s test was then performed to evaluate publication bias by comparison of P values, which if <0.10 indicating significant publication bias.Initially, we performed standard pairwise meta-analysis with a random-effects model by using STATA 12.0 . For indirect and mixed comparisons, a Bayesian network meta-analysis followed, with a random-effect hierarchical model by means of Markov chain Monte Carlo methods with Gibbs sampling from 1,000 iterations obtained after a 5,000 iteration training phase. We performed random-effects pairwise and network meta-analyses to obtain estimates for the incidence of HCC. According to Cochrane Handbook for Systematic Reviews of Interventions Version 5.1.0, the incidence of HCC was presented as RR with 95% CIs. Clinical heterogeneity was assessed with the IP value.To rank the treatments for an outcome, we estimated the relative ranking probability of each treatment and obtained the treatment hierarchy of competing interventions using rankograms, surface under the cumulative ranking probabilities. To check for the presence of inconsistency, we used the previously described node-splitting method, which separates evidence for a particular comparison into direct and indirectHow to cite this article: Zhou, Y.-Y. et al. Systematic Review with Network Meta-Analysis: Antidiabetic Medication and Risk of Hepatocellular Carcinoma. Sci. Rep.6, 33743; doi: 10.1038/srep33743 (2016).


CD24) Ala57Val polymorphism has been implicated as a risk factor for multiple sclerosis (MS) and systemic lupus erythematosus (SLE); however, genetic studies have produced controversial results. A meta-analysis was performed on this topic. We used odds ratio (OR) and 95% confidence interval (95% CI) to investigate the strength of association. Eleven studies from nine publications consisting of 2466 cases and 2650 controls were included. The results suggested that the CD24 Val/Val genotypes were associated with an increased risk of MS in all study subjects and Caucasians . Sensitivity analysis showed that no individual study was found to be significantly biasing the pooled results. Although meta-analysis also suggested an association between the CD24 Val/Val genotypes and SLE risk in Caucasians , sensitivity analysis demonstrated that the association was not statistically significant after removing a Spanish study. In conclusion, this meta-analysis suggests that the CD24 Ala57Val polymorphism is associated with an increased risk of MS in Caucasians. However, the available evidence is not sufficient to support an association between the CD24 Ala57Val polymorphism and SLE risk.The cluster of differentiation 24 ( HLA) gene on chromosome 6p21 is the most important genetic factor for MS and SLEHLA-DR15 haplotype (DRB1*1501-DQA1*0102-DQB1*0602) and its individual alleles appear to have the main role in CaucasiansHLA-DR2 and DR3 alleles and the Class III variants are important genetic susceptibility factors for the disease6Multiple sclerosis (MS) and systemic lupus erythematosus (SLE) are autoimmune diseases mediated by self-reactive T cells and other cells of the adaptive and innate immune systems1CD24 gene, which maps on 6q21, is considered a candidate gene for MS and SLE. A C > T single-nucleotide polymorphism (SNP) is located in the presumable GPI-anchor cleavage site of the CD24 protein (rs52812045), leading to the replacement of an alanine (CD24 Ala) by a valine  CD24 Ala57Val polymorphism with MS and SLE17181921222324CD24 Ala57Val polymorphism and risk for MS and SLE.Cluster of differentiation 24 (CD24) is a small cell surface protein molecule anchored by glycosyl-phosphotidyl-inositol (GPI) in a wide variety of cell types, including T cells, B cells, dendritic cells (DCs), cancer cells and local antigen-presenting cells in the CNS891011in rs528145, leadiCD24 Ala57Val polymorphism with MS and SLE. PubMed, Scopus, Web of Knowledge, China National Knowledge Infrastructure (CNKI) and Wanfang databases were searched for papers published until January 2015 without language restrictions. Search terms included “multiple sclerosis”, “MS”, “systemic lupus erythematosus”, “SLE”, “cluster of differentiation 24”, “CD24”, “genetics”, “polymorphism”, “SNP”, “rs52812045”, and “association”. The reference lists of retrieved publications were also reviewed to identify other relevant studies missed by the database search. The following inclusion criteria were used for study selection: (1) case-control design; (2) genotype frequency data for both cases and controls were available; (3) literature published in English or Chinese. Studies were excluded if they met the following criteria: (1) studies on animal populations; (2) no control subjects; (3) duplicate data; (4) insufficient data. Reviewers screened study reports by first screening titles and abstracts to select publications for full-text evaluation, then screening full-text publications to confirm eligibility of the studies.An electronic search of the medical literature was conducted to identify genetic association studies assessing the relationship of the Data were extracted by two of the authors, and differences were resolved by consensus. For each study, the following data were extracted: first author name, publication year, country, ethnicity, disease, sample size, genotypic frequencies of the CD24 Ala57Val SNP, diagnostic criteria and genotyping method. We assumed study subjects to be Caucasians, if Caucasians comprised 90% of the subjects. When a publication reported results on different subpopulations according to country, each subpopulation was considered as a separate study in this meta-analysis.www.ohri.ca/programs/clinical_epidemiology/oxford.asp), awarding points based on selection of cases and controls, comparability of cases and controls, and ascertainment of exposure  , Ala/Val versus Ala/Ala (OR2), and Val/Val versus Ala/Val (OR3). The resulting ORs were used to determine the most appropriate genetic model using a previously described approachIf OR1 = OR3 ≠ 1 and OR2 = 1, a recessive model is suggested;If OR1 = OR2 ≠ 1 and OR3 = 1, a dominant model is suggested;If OR2 = 1/OR3 ≠ 1 and OR1 = 1, an over dominant model is suggested;If OR1 > OR2 > 1 and OR1 > OR3 > 1 (or OR1 < OR2 < 1 and OR1 < OR3 < 1), a co-dominant model is suggested;Meta-analysis was conducted using STATA version 11.0. Odds ratios (ORs) with 95% confidence intervals (CI) were used to evaluate the size and strength of relationship of the CD24 Ala57Val polymorphism with MS and SLE. ORs were calculated from genotype frequency data without adjustment. Z-test was used for assessing the significance of the pooled OR, with P-value of <0.10 for the Q test was considered to indicate heterogeneity across studies, in which case the DerSimonian Laird random-effects model was fitted to calculate the pooled ORP < 0.05 being considered statistically significant. The pooled frequency of the Val allele in control groups from Caucasian studies was calculated using Meta-Analyst version 3.13. A power calculation was carried out using Power and Sample Size Calculation version 3.1.2 (http://biostat.mc.vanderbilt.edu/twiki/bin/view/Main/PowerSampleSize). To assess whether the genotype frequencies in control groups were in Hardy-Weinberg equilibrium (HWE) we used a web-based programme .Heterogeneity was assessed by Cochrane's Q test and the I-square statistic. A CD24 Ala57Val polymorphism and MS risk1718192123202217192021222318A flow diagram showing the selection process of studies included in the present analysis is shown in 1718192021222324zP < 0.001), 1.17 , and 2.09 , respectively. These genotype-specific ORs were most suggestive of a recessive model . In this model, meta-analysis demonstrated a significant association between the CD24 Ala57Val polymorphism and MS risk in all study subjects   . Sensitivity analysis was conducted to confirm the stability and liability of the meta-analysis by removing each of the involved study in turn. No individual study was found to be significantly biasing the pooled results (data not shown). A cumulative meta-analysis based on the publication date showed a significant and consistent trend toward increased risk of MS for the Val/Val genotypes over time and the trend stabilized by 2009   . Subgrou< 0.001) . Ethnici by 2009 . Heterogctively) . When wezP < 0.001), 1.17 , and 1.54 , respectively. These estimates suggested a recessive effect of the T allele  for SLE. Therefore, Val/Val was compared with Ala/Val and Ala/Ala genotypes combined. In this model, a significant association between the CD24 Ala57Val polymorphism and SLE risk was found in all study subjects   . We further performed subgroup analysis in Caucasian studies according to sample size (subjects ≥ 1000 or subjects < 1000) and genotyping method. In subgroup analysis based on sample size, the I2 value for heterogeneity decreased from 51.2% to 0%  . A cumul< 0.001) . Subgrou< 0.001) . However2% to 0% , but it CD24 Ala57Val polymorphism with MS and SLE in recessive model (Funnel plots offer a visual sense of the relationship between effect size and precision for publication bias amongst the studies in meta-analysis. ve model  for SLE.ve model .CD24 Ala57Val polymorphism has been implicated in the susceptibility to MS and SLE and has been assessed in many association studies. However, the outcomes of these studies remain contradictory rather than convincing. The inconsistency seems to be mainly owing to the heterogeneity of the populations under study and the small sample size of most studies, leading to little statistical power. Meta-analysis is a quantitative statistical analysis that integrates the results of several independent studies in order to see whether the overall effect is significant. It has been increasingly used in evaluating the association between predisposing genes and complex traits, including MS, SLE, type 1 diabetes and inflammatory bowel diseaseCD24 Ala57Val polymorphism with MS and SLE. The results indicated that the CD24 Val/Val genotypes were associated with increased risk of MS in Caucasians, whereas current evidence is still insufficient to suggest an association between the CD24 Val/Val genotypes and SLE risk.It is generally believed that variants of immune-related genes modulate genetic predisposition to autoimmune diseases, including MS and SLE. The CD24 Ala57Val polymorphism and MS, two studies presented deviation from HWE. After exclusion of them, the association in all study subjects and Caucasians remained statistically significant. In addition, sensitivity analysis and cumulative meta-analysis further strengthened the validity of the results. Thus, the association of the CD24 Val/Val genotypes with MS revealed by this meta-analysis can be considered reliable. Although meta-analysis also showed an association between the polymorphism and SLE in Caucasians, sensitivity analysis demonstrated that the association was not statistically significant after removing a Spanish cohortCD24 Ala57Val polymorphism and SLE in additional ethnic groups to achieve a definitive conclusion.Among eligible studies evaluating the In this meta-analysis, low or moderate heterogeneity was identified. Subgroup analyses indicated that diagnostic criteria was the major factor for heterogeneity among Caucasian studies evaluating the polymorphism and MS, whereas sample size was largely responsible for the heterogeneity of results of SLE.CD24-deficiency leads to a reduction in late pre-B and immature B-cell populations in the bone marrow, suggesting that CD24 expression influences the maturation of B cellsCD24 gene may be a candidate gene for MS and SLE. It has been speculated that the CD24 Ala57Val SNP leads to a nonconservative replacement from alanine to valine at the site which immediately precedes the putative cleavage site for GPI anchorCD24 Val/Val genotypes. Zhou et al. reported that CD24 Val allele lead to 30–40% more cell surface expression of CD24 than the CD24 Ala alleleCD24 Val allele was associated with the production of autoantibodies in patients with SLECD24 SNP in MS and SLE, more functional assays are needed in the future.CD24 is expressed in various immune cell types and is known to mediate several important functions. Previous work has established that CD24 functions as a costimulatory regulator for T cell clonal expansion. CD24 expression on T cells is required for the homeostatic proliferation of both CD4 and CD8 T cells in the lymphopenic environmentCD24 Ala57Val polymorphism and risk of MS and SLE may not have been reported in the literature and therefore could not be included in our meta-analysis. Second, the results of our meta-analysis were based on the data from case-control studies, because they were most frequently used in association studies to evaluate the CD24 Ala57Val polymorphism. We expect that as more studies of other types, such as prospective cohort and family-based studies become available, a more comprehensive systematic review and meta-analysis can be conducted to provide additional information on the role of this SNP in MS and SLE. Third, we only evaluated the CD24 Ala57Val polymorphism in this meta-analysis. Because of the limitation of the published data, we were unable to assess the relationship of other CD24 polymorphisms, such as the −809 C/A, −534A/C and −492G/C polymorphisms in the promoter region with MS and SLE. There were insufficient data to evaluate linkage disequilibrium between the CD24 Ala57Val SNP and other polymorphisms. Fourth, because most eligible studies were conducted in Caucasians, the results of this meta-analysis were mainly applicable to this ethnic group. Subgroup analysis could not be performed in Asians due to insufficient data. Future studies should be conducted in other ethnic groups besides Caucasians.The present meta-analysis has some limitations that should be acknowledged. First, the key threat to any meta-analysis is that of reporting bias. Although there is no evidence of publication bias , it is nCD24 Ala57Val polymorphism are associated with increased risk of MS in Caucasians. However, the available evidence is not sufficient to support an association between the CD24 Ala57Val polymorphism and SLE risk.In conclusion, the present meta-analysis of the literature reveals that the Val/Val genotypes of the Supplementary Information


It is still inconclusive whether alcohol consumption affects the risk of thyroid cancer. We conducted a meta-analysis of available epidemiological data to address this issue.P = 0.112).Compared with nondrinkers, the pooled relative risks (RRs) and corresponding 95% confidential intervals (CIs) of thyroid cancer were 0.80 (95% CI 0.71-0.90) for any drinkers, 0.81 (95% CI 0.70-0.93) for light and 0.71 (95% CI 0.63-0.79) for moderate drinkers. The dose–response analysis suggested that there is no evidence of a dose-risk relationship between alcohol intaking and thyroid cancer risk (Eligible studies were identified by searching PubMed and EMbase databases. A total of 24 studies, included 9,990 cases with thyroid cancer, were included in this meta-analysis. We defined light alcohol intake as ≤ one drink/day and moderate as >one drink/day. The summary risk estimates were calculated by the random effects model. A dose-response analysis was also conducted for modeling the dose-risk relation.This meta-analysis confirmed an inverse association between alcohol consumption and thyroid cancer risk. Further studies are needed to better understand the potential mechanisms underlying this association. Thyroid cancer is now recognized as a major public health issue worldwide because of its rapidly increased incidence –3. HowevThe literature search strategy identified 154 potentially eligible reports. Finally, 24 studies met the inclusion criteria, including 7 cohort studies , 32, 33 versus nondrinkers). The pooled RR, based on overall studies, was 0.80 (95% CI 0.71-0.90). The corresponding estimates were 0.87 (95%CI: 0.78-0.96) for cohort studies and 0.75 (95%CI: 0.63-0.89) for case-control studies. Satisfactory homogeneity was found among cohort studies . However, there were significant heterogeneity among case-control studies .Figure Figure P for heterogeneity < 0.01). Thirteen studies reported data for women, with a pooled RR of 0.81 , while only five studies reported data for men, with a pooled RR of 0.77 . Considering the geographical area, the pooled RR for thyroid cancer risk was 0.77 (95%CI: 0.65-0.92) in European studies, 0.79 (95%CI: 0.67-0.93) in American studies and 1.01 (95%CI: 0.58-1.75) in Asian studies. Most studies provided multivariate-adjusted results. The adjusted and unadjusted RRs were 0.82 (95%CI: 0.72-0.93) and 0.73 (95%CI: 0.57-0.95) for thyroid cancer risk. Significant heterogeneity was found within the geographical area and whether adjusted for multivariate subgroups .Table P < 0.10), which may be partly sources of heterogeneity for overall results. However, all of these factors didn't have statistically significant association with the sources of heterogeneity in univariate or multivariate meta-regression analysis . Table P = 0.112). There is no linear relationship for thyroid cancer with every 12.5 g/d alcohol increase .As shown in Figure In this systematic review, we quantitatively summarized all the available epidemiological evidence on the association between alcohol consumption and thyroid cancer risk and found that any drinking, in relation to nondrinkers, was associated with a 20%, light drinking (≤12.5 g/day of ethanol) with a 19%, and moderate drinking (> 12.5 g/day of ethanol) with a 29% reduced risk for thyroid cancer in present study. However, the association of heavy drinking (> 50 g/day of ethanol) and thyroid cancer risk was unclear due to the lack of quantitative data provided from original studies. The dose-response analysis didn't show a significant dose-response association between alcohol drinking and thyroid cancer risk, but the dose-risk curve, as shown in Figure versus 0.75). However, among case-control studies, the evidence for an inverse association between alcohol consumption and thyroid cancer risk was limited to the population-based studies. Hospital-based case-control studies did not show a statistical significant association. This may be explained by a higher risk of inappropriate control selection, less control for confounding by smoking or other confounders in hospital-based studies as compared with population-based studies  and American studies , but not in Asian studies . This might be due to the different alcohol beverages, different lifestyles and the difference of functional variants in gene involved in alcohol metabolism, such as single nucleotide polymorphisms in ADH1B and ALDH2, among various ethnic groups . The genSmoking has been suggested as a protective factor for thyroid cancer by several studies , 53, whiThere are several potential explanations for the effect of alcohol intake on decreased risk of thyroid cancer. First, light-to-moderate alcohol consumption is associated with enhanced insulin sensitivity and reduced type 2 diabetes –56. DiabThere are some potential limitations to our study. First, marked heterogeneity was detected in several analyses which may reflect differences in study design, study population and adjustment for confounders. Nevertheless, the random-effects models were used to take heterogeneity into account. Meanwhile, subgroup analysis, sensitivity analysis and meta-regression analysis were used to explore the potential heterogeneity in our meta-analysis. Second, former drinkers and sick quitters might have been misclassified into the reference group in some individual studies and thus a decreased effect has been estimated. We could not address this issue because the majority of the studies did not report separate estimates between former drinkers and lifelong never drinkers. Third, most studies (13/24) collected information by self-reporting questionnaires, which might lead to information bias. However, several studies have found satisfactory correlation coefficients for validity and reproducibility of self-reported alcohol drinking in various populations .Notwithstanding the limitations discussed above, this meta-analysis includes the most comprehensive information with a large sample size on alcohol consumption and thyroid cancer risk to date, and confirms an inverse association of alcohol consumption for thyroid cancer risk. Further studies are needed to better understand the potential mechanisms underlying this association.We identified all relevant case-control and cohort studies published in English by searching the two databases (PubMed and EMbase) from the beginning of indexing to August 2015, using the following terms:  AND (thyroid tumor OR thyroid cancer OR thyroid carcinoma OR thyroid neoplasm) AND (cohort OR prospective OR case-control), following the Meta-analysis Of Observational Studies in Epidemiology (MOOSE) guidelines . Two autTwo authors (Wang XF and WL Cheng) independently extracted data from each original article using a pretested form. The extracted data were composed of first author, year of publication, country, gender, study design, period of enrollment and source of controls (case-control studies), follow-up time (cohort studies), sample size , variables adjusted for estimates, risk estimates  for alcohol consumption levels and the corresponding 95% CIs. All risk estimates results were expressed as RRs. The ORs and HRs approximate to the RRs because of the low incidence of thyroid cancer in general population.Two authors (Wang XF and Li JD) independently assessed the quality of included studies according to the Newcastle-Ottawa Scale (NOS) . The NOSThe expression of alcohol intake was various in different studies; therefore, we used grams of ethanol per day as a standard measure and defined one drink as 12.5 g of ethanol, 1 ml of alcohol as 0.8 g and 1 ounce as 28 g. Nondrinkers drinkers were defined as the reference category. Since the highest category of alcohol exposure was set at > 1 drinks/day in the majority of studies, thus we divided the alcohol drinker into two levels: light drinker defined as ≤1 drink/day (≤12.5 g/day of ethanol) and moderate as >1 drinks/day (>12.5 g/day of ethanol). When a study provided a series of categories of alcohol consumption, we used the midpoint of each category as the corresponding exposure dose. For the upper open-ended category, the exposure dose was calculated as 1.2 times its lower bound . When thI2 statistic [I2 more than 50% or Q-test reporting a P value < 0.1. To explore the potential sources of heterogeneity among studies, we conducted meta-regression analysis and subgroup analyses in strata of gender, study design, control source for case-control study, geographic area, publication year, number of cases, tumor subtype, NOS quality score and whether adjusted for multivariate. Sensitivity analyses were also performed by excluding each study at a time to clarify the influence of each study on the overall estimates. Publication bias was assessed by Begg's rank correlation test [The multivariate-adjusted RRs were used for the meta-analyses; when they are unavailable, the unadjusted RRs were calculated from original data. The pooled RRs with 95% CIs were calculated by random effects models  for the tatistic . A signiion test  and the ion test .A potential nonlinear dose-response relationship between alcohol consumption and thyroid cancer risk were conducted based on the natural logarithm of the RR for each study with at least three quantitative categories of exposure using the methods described by Orisini . The P vP value less than 0.05 was considered to be statistically significant. All P values were two-tailed. All calculations were performed using Stata 11.0 .A


Lung cancer, as the leading cause of cancer mortality worldwide, has been linked to environmental factors, such as air pollution. Residential exposure to petrochemicals is considered a possible cause of lung cancer for the nearby population, but results are inconsistent across previous studies. Therefore, we performed a meta-analysis to estimate the pooled risk and to identify possible factors leading to the heterogeneity among studies.The standard process of selecting studies followed the Cochrane meta-analysis guideline of identification, screening, eligibility, and inclusion. We assessed the quality of selected studies using the Newcastle-Ottawa scale. Reported point estimates and 95% confidence intervals were extracted or calculated to estimate the pooled risk. Air quality standards were summarized and treated as a surrogate of exposure to air pollution in the studied countries. Funnel plots, Begg’s test and Egger’s test were conducted to diagnose publication bias. Meta-regressions were performed to identify explanatory variables of heterogeneity across studies.I2 = 25.3%). Such effect was stronger by a factor of 12.6% for the year of follow-up started 1 year earlier .A total of 2,017,365 people living nearby petrochemical industrial complexes (PICs) from 13 independent studied population were included in the analysis. The pooled risk of lung cancer mortality for residents living nearby PICs was 1.03-fold higher than people living in non-PIC areas (95% CI = 0.98–1.09), with a low heterogeneity among studies  contains supplementary material, which is available to authorized users. Lung cancer is the leading cause of cancer deaths globally . The GloPetrochemical manufacturing industry, defined as petroleum refining  or industrial organic chemicals manufacturing SIC 2869), involves processes that produce and potentially emit hazardous chemicals into the surrounding air, soil, and water. These petrochemical manufacturing factories are usually clustered in an industrial area together with other manufacturing processes or industry, such as steel, coking, and thermoelectric plants , 8, and 69, involSeveral epidemiological studies have explored associations between the PICs and lung cancer risks of nearby people. Given high public concerns of health, the US started several investigations of suspected cancer risks for people living nearby PICs back to the 1970s –17. For Due to the inconsistent results, our study aimed to estimate lung cancer mortality risk associated with the PICs by combining cross-country data from different studies via a systematic review and meta-analysis.Lung cancer OR Lung neoplasm) AND " as the search term. Two researchers—HY Hung and RT Lin—selected independently articles that met the inclusion and exclusion criteria as below.We selected exclusively articles from PubMed, Cochrane Library, Web of Science, Science Direct, and other sources that published before July 11, 2017. We used "(The inclusion criteria were: (1) original articles that clearly defined exposure group as residents living nearby PICs; (2) original articles that clearly defined lung cancer mortality according to International Classification of Diseases (ICD); (3) original articles that reported either confidence intervals (CI), standard errors (SE), or both; and (4) original articles that were written in English and full-texts were available. The exclusion criteria were: (1) studies with subjects overlapped with other publication; (2) studies that focused on occupational exposure in petrochemical plants only; and (3) studies that reported lung cancer incidence only and lack of mortality data.Figure −3 per year [Since lung cancer mortalities were less than 10per year , we coulper year , 26. Theper year , 27.lnRR using following equations:For the study not reporting CI or SE , we estilnRR) represents the variance of natural log of relative risks (RRs); R1 and R0 represents mortality rates of the studied group and the reference group, respectively; and SE(lnRR) represents the standard error of natural log of relative risks.where Var, and bona fide observation (defined as 10 or more years of observation after 20 years of PIC operations vs. less than 10 years). Then, we applied meta-regressions to investigate the possible factors of heterogeneity, including sex, ethnicity, location, year of publication, and the starting year of follow-up. We also conducted sensitivity analyses to assess the influence of individual study on the overall RR by adding one estimate into the pooled estimates gradually. Finally, we used a funnel plot and the Begg’s and Egger’s regressions for asymmetry test to examine whether there was publication bias and small-study bias. All analyses were performed using the Stata Software version 11.2 . We set the statistical significance level at 0.05, using a two-sided test.We applied a random-effects model to examine whether there were within- and between-study heterogeneities using the I2 test . We set handbook . We furtTo assess the risk of bias in each study, the quality of each study was recorded and assessed using the Newcastle-Ottawa Quality Assessment Scale . Records10, and PM2.5 [Pollutants emitted from PIC might vary over time, likely due to the change of manufacturing process and pollution control technology. Since data on air quality around studied petrochemical areas were limited, we reviewed national or regional ambient air quality standards in studied countries or regions: European Union (EU), Taiwan, and the US. We summarized three air quality standards for studied countries, including total suspended particles (TSP), PMnd PM2.5 –41.Table I2 was 25.3%, indicating low heterogeneity existed among these studies.Figure 10 standard, latency period, and bona fide observation. For each characteristic, there was no significant difference among pooled estimates between subgroups based on overlapping 95% CIs. However, we found a higher risk of lung cancer associated with residential exposure to PICs in the era of looser PM10 standard .Table p-value = 0.034, Fig. Except for the starting year of follow-up, we did not find any possible heterogeneous factor from the meta-regression analysis. The slope of the meta-regression line suggested that for an increment in the starting year of follow-up, the RR of lung cancer would be 0.874-fold lower . The funnel plot also indicated no asymmetry for the estimates for the 13 study groups was observed . Stratification analysis by different characteristics, such as sex and ethnicity, did not change the magnitude of this association. In contrast, the starting year of follow-up affected the association between lung cancer mortality and exposure to PICs by a factor of 0.874. That is, the estimated risk of lung cancer mortality was higher among subjects recruited in earlier periods, and the risk decreased by 12.6% if the year of follow-up started 1 year later.n = 2,017,365) and diverse populations  made the pooled estimate more representative and enhanced the generalizability. Third, by applying the random-effect model, we were able to address the heterogeneity between studies and further reported the pooled effects.The scientific evidence of the study is sound and solid from several perspectives. First, the outcome variable was based on pathological samples and/or the ICD-9. Individual data were obtained by linking to governmental database. Second, the large sample size Additional file 2:Data quality assessment on the Newcastle-Ottawa Quality Scale. (DOCX 15 kb)


VEGF gene polymorphisms can alter the protein concentrations and might be associated with renal cell carcinoma risk. However, the results of studies investigating the association between VEGF polymorphisms and renal cell carcinoma risk are inconsistent. Thus, a meta-analysis was performed.Vascular endothelial growth factor (VEGF) protein plays important role in renal cell carcinoma (RCC) development and progression. We selected eligible studies via electronic searches. Only high-quality studies were included based on specific inclusion criteria and the Newcastle-Ottawa Scale (NOS).Eight studies primarily focusing on seven polymorphisms were included in our meta-analysis. Our results showed dramatically high risks for renal cell carcinoma were found regarding most genetic models and alleles of the +936C/T polymorphism (except CT vs. CC). In addition, significant increased renal cell carcinoma risks were found regarding all genetic models and alleles of the -2578C/A polymorphism. However, no significant associations were found between renal cell carcinoma risk and the +1612G/A, -460T/C, -634G/C, -405G/C or -1154G/A polymorphisms.VEGF are associated with an increased risk for renal cell carcinoma. Additional rigorous analytical studies are needed to confirm our results.Our meta-analysis indicates that the +936C/T and -2578C/A polymorphisms of  Approximately 337,860 cases of renal cell carcinoma (RCC) are diagnosed annually, and nearly 143,406 patients die from this cancer each year worldwide . The etiVEGF signaling suppresses both tumor-induced angiogenesis and tumor growth [VEGF gene is located at chromosome 6p21.3 and consists of 8 exons. At least 30 single nucleotide polymorphisms (SNPs) exist in this gene [Vascular endothelial growth factor (VEGF) is an important pro-angiogenic growth factor, and it is one of the most potent endothelial cell mitogens , 11. VEGr growth . The VEGhis gene  and somehis gene .VEGF polymorphisms and RCC risk in diverse populations; however, the results of these studies conflict. To examine the association between VEGF polymorphisms and RCC risk, we performed a meta-analysis of all eligible published data up to June 5, 2016.Recently, numerous studies have been performed to evaluate the association between VEGF polymorphisms and RCC risk. After the subsequent data extraction, one study was excluded because it lacked controls [VEGF polymorphisms and RCC risk  , 20, 21,We also investigated the -405G/C (rs2010963) , 19 and Hardy-Weinberg disequilibrium was observed in two studies .Except for the -405G/C rs2010963) and -1154G/A (rs1570360) polymorphisms, we used both funnel plots and Egger's test to assess the publication bias of each genetic model and allele. Our results did not show a publication bias for most of the genetic models and alleles  polymorphism is located in the 3′-UTR and likely associated with obviously increased serum VEGF levels [Zhang et al.  previousF levels , which aF levels , the carF levels . FurtherF levels , 35. HowCertain limitations of this meta-analysis should be acknowledged. First, because our study only considered published articles, a publication bias might exist. However, the publication bias was only found for the CC vs. CT+TT of -460T/C (rs833061) polymorphism. The statistical results of the funnel plot and Egger's test support this finding. Second, the heterogeneities among certain genetic models and alleles were significant. The reasons underlying these heterogeneities included the source of the controls, the study design and differences in genetic backgrounds. Third, the control sample of two articles were in Hardy-Weinberg disequilibrium, however, all the results of +936C/T (rs3025039) and -2578C/A (rs699947) polymorphisms did not change significantly after sensitivity analyses. Fourth, as the most of the cases of +936C/T and -2578C/A polymorphisms were from Asians, so our results of these two SNPs may not represent Caucasians. Finally, because of the use of unadjusted data, potential confounds such as age, sex and residence might also have affected the effect estimates. Thus, a more precise and large scale evaluation based on adjusted data is needed.VEGF are associated with increased risks for RCC. However, no significant RCC risks were obtained regarding the +1612G/A (rs10434), -460T/C (rs833061), -634G/C (rs2010963), -405G/C (rs2010963) or -1154G/A (rs1570360) polymorphisms. To the best of our knowledge, this meta-analysis is the first to report that the +936C/T (rs3025039) polymorphism can increase the risk of RCC. Larger and more rigorous analytical studies are required to confirm our results and evaluate the gene-environment interactions with regard to RCC risk.In summary, our meta-analysis suggests that the +936C/T (rs3025039) and -2578C/A (rs699947) polymorphisms of VEGF polymorphisms and the risk of RCC in humans; (2) used a case control design; (3) confirmed RCC using the accepted diagnostic criteria; (4) reported sufficient published data, including ORs and their 95% CIs, or the number of events for the purposes of calculation. The exclusion criteria were (1) a lack of sufficient data to calculate ORs with corresponding 95% CIs; and (2) overlapping cases or controls. Only the most recent or the largest research study was included in the case of overlap.According to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA), we performed an electronic systematic search of PubMed, the Cochrane Library database, EMBASE, Google Scholar and the China National Knowledge Infrastructure (CNKI) without any restriction on language up to June 5, 2016. The combinations of keywords used were “renal cancer” or “renal carcinoma”; “polymorphism” or “variant”; and “vascular endothelial growth factor” or “VEGF.” In addition, the reference lists of the papers retrieved and recent reviews were also examined. We included all studies that (1) evaluated the association between P-value. All disagreements were resolved via discussion.Two investigators (GMC and DWJ) extracted the raw data independently based on the inclusion and exclusion criteria. The following information was extracted from all of the enrolled studies (see Table Two authors (GMC and SZR) assessed the study quality using the NOS  which evVEGF polymorphisms and the risk of RCC was evaluated via pooled ORs with 95% CIs. The significance of the pooled ORs was tested using the Z-test, and a (two-tailed) P-value of <0.05 was regarded as significant. The HWE was calculated in the control groups using the chi-square test, and P<0.05 signified a departure from HWE. Between-study heterogeneity was calculated using the 2I test. If the heterogeneity was significant (2I>50%) [P<0.05 was considered as significant publication bias) [The relationship between (I2>50%) , then a (I2>50%) ; otherwi(I2>50%)  was applon bias) . All sta


The evidence for association between Epstein-Barr virus (EBV) infection and risk of oral squamous cell carcinoma (OSCC) is inconsistent in the literature. Therefore, this meta-analysis was conducted to clarify this association.odds ratio (OR) and 95% confidence interval (95% CI) were estimated to determine the association between EBV infection and OSCC risk using a fixed- or random-effects model based on heterogeneity. Publication bias was assessed using funnel plot analysis.A literature search was conducted in electronic databases for English- and Chinese-language publications until March 31, 2017 to include eligible case-control studies. The pooled OR with 95% CI between EBV infection and OSCC risk was 5.03 (1.80–14.01) with significant heterogeneity observed (I2 = 87%). The subgroup analysis indicates that the year of publication, study location, economic level, sample size, tissue type, detection method and marker, control type, and language might explain potential sources of heterogeneity. Publication bias was not observed, and sensitivity analysis showed stable results.A total of 13 case-control studies with 686 OSCC patients and 433 controls were included based on predetermined inclusion and exclusion criteria. The pooled The results of the current meta-analysis suggest that EBV infection is statistically associated with increased risk of OSCC. However, additional high-quality studies with larger sample sizes are needed to further confirm the relationship between EBV and OSCC. Oral squamous cell carcinoma (OSCC) is the most common subset (90%) of oral cancer with a global incidence of 275,000 cases annually , the sixEBV is an oncogenic human herpes virus that contains double-stranded DNA, known as the first human tumor virus ,8. It apThe strength and consistency of EBV DNA present in OSCC indicate a potentially important role of EBV infection on OSCC pathogenesis. However, controversial results have been reported –13. TherThe current meta-analysis was conducted based on the Meta-Analysis of Observational Studies in Epidemiology guidelines . Pubmed,in situ hybridization (ISH), and immunohistochemistry (IHC); and (7) full text available in English or Chinese. Case series, reports, animal models, in vitro studies, reviews, editorials, conference abstracts, and letters without sufficient data were excluded. If multiple publications reported results based on the same study, the most recent article or the article with bigger sample size was included.The titles and abstracts of all relevant studies were independently examined by two authors (YYS and MZ) to assess eligibility. Studies were included based on the following criteria: (1) evaluated the association between EBV infection and patients with OSCC through the expression of EBV level  in tissue samples; (2) confirmed histopathological diagnosis of OSCC cases; (3) reported original data; (4) used a case-control study design; (5) employed fresh, frozen, or paraffin-embedded (PE) storage methods; (6) used polymerase chain reaction (PCR), reverse transcription PCR (RT-PCR), real-time quantitative PCR (qPCR), The data extraction was performed by two authors (YYS and MZ) independently using a pre-designed data extraction form based on the guidelines for meta-analysis . DiscrepThe Newcastle-Ottawa Scale (NOS) was used to evaluate the methodological quality of the included studies , where fodds ratios (ORs) and the corresponding 95% confidence intervals (CIs) based on weighted pooled measures. Forest plots were generated to visually evaluate the study-specific and pooled effects. Heterogeneity across studies was evaluated using the Cochran’s Q test with a significant level of P <0.10 or the I2 statistic >50% [P <0.10 or I2 >50%. Otherwise, the fixed-effects model was used [P values were two-sided, with P <0.05 considered statistically significant, except where otherwise specified.The numbers of OSCC cases and controls were used to calculate the pooled tic >50% ,17. The was used . Sensitiwas used . Begg’s was used ,21. Funnwas used . All staA total of 678 citations were retrieved from Pubmed, Web of Science, Cochrane, Embase, CNKI, Wanfang data, CQVIP, and CBM. After excluding 153 citations due to duplication, 525 unique citations were considered further. Of these, 500 citations were sequentially excluded after the first screening based on the abstracts or titles . The fulOR 5.03   to 6.53   . Sensiti5–17.40) .ORs (95% CIs) for the association of EBV infection and OSCC from different study locations were 4.17 (0.69–25.21), 9.69 (3.16–29.72), and 1.96 (0.24–15.91) in Asia, Europe and USA, and Africa, respectively; 3.83 (0.94–15.53) and 6.92 (1.35–35.40) for developing and developed countries, respectively; 2.17 (0.44–10.80) and 10.47 (2.39–45.79) for sample sizes <100 and ≥100, respectively; 4.25 (0.99–18.29) and 6.17 (1.27–29.95) for studies published before and on or after 2000, respectively; 2.39 (0.78–7.33) and 31.27 (5.79–169.01) for PE and fresh/frozen tissues, respectively; 3.37 (0.90–12.65) and 7.62 (2.50–23.22) for studies applying PCR and Nested PCR, respectively; 4.36 (1.42–13.40) and 16.76 (0.22–1276.16) for studies detecting EBV DNA, and EBV RNA and protein, respectively; 4.78 (1.60–14.33), 10.37 (5.30–20.28), and 4.37 (1.16–16.97) for studies randomly selected, age-matched, and neither random nor age-matched control samples, respectively; and 6.29 (2.49–15.89), and 4.84 (1.48–15.88) for Chinese- and English-language publications, respectively. The meta-regression analysis revealed that the selected study characteristics did not significantly influence the results (P >0.05), except in the type of tissues (P = 0.03), and only partially explained the source of heterogeneity.To explore the heterogeneity of the study, subgroup analyses and meta-regression on a number of key study characteristics were performed . EBV infCI, 0.28–19.56) and 5.02  cumulative estimates; however, the association between EBV infection and OSCC risk was relative weaker in the earlier than the more recent studies with a higher cumulative estimate and narrower 95% CI.A cumulative meta-analysis was conducted to evaluate the temporal trend of the pooled results using random-effects model . Most cuP = 0.222; Egger’s test, intercept = 3.45, t = 1.98, and P = 0.073). Visual inspection of the Begg's funnel plot revealed a nearly symmetrical distribution, confirming the absence of publication bias  and viral genome DNA, were used to detect EBV ,29,31,32The results of this study indicated that EBV DNA, mRNAs, and EBV proteins were expressed in the majority of OSCC cells ,28. Two P = 0.03). However, the use of PE tissues promotes easier and simpler DNA detection. Indeed, according to a previous study, Deacon et al. [Based on our results, the fresh/frozen tissues showed a slightly higher EBV positivity rate (49.09%) than that of PE tissues (48.18%). The results of meta-regression indicate that different methods of sample storage might contribute to heterogeneity of the results Click here for additional data file.S1 Data(XLSX)Click here for additional data file.


P < 0.001). Additional studies with larger patient cohorts are required to validate our findings.We investigated the relationship between eczema and the risk of primary glioma. Relevant studies were selected through electronic searches of PubMed and EMBASE. A meta-analysis of 12 case-control studies and one cohort study was performed. A fixed effect model was applied to analyze 13 studies consisting of 10,897 glioma cases and 56,081 controls. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated to assess the strength of the associations. The data demonstrate that eczema significantly reduces the risk of glioma (OR = 0.69, 95% CI = 0.61–0.78,  Glioma is the most common type of brain tumor worldwide. It affects approximately six per 100,000 individuals annually . The proAtopic dermatitis, or eczema, is common skin condition worldwide . PatientOnly 0.05%–4.7% of glioma patients survive for 5 years after diagnosis . PreviouThe literature selection process is shown in Figure The author information, publication years, numbers of control and glioma cases, odds ratios (ORs), and 95% confidence intervals (CIs) for glioma in individuals with eczema for the studies that were included in the analysis are shown in Table 2 = 22.25, 2I =46.1 %, P = 0.035). Therefore, a fixed-effects model was selected in order to calculate the ORs. In the pooled analysis, individuals with a history of eczema had a decreased risk of incident glioma .An assessment of heterogeneity suggested that the 13 studies included in the analysis were homogeneous  , 37, 38.P < 0.001), suggesting that eczema may modify the risk of glioma through modulating the immune system. Our results are consistent with those of previous studies, which demonstrated an association between allergies and glioma [P < 0.001) based on a meta-analysis of 12 observational studies. However, the sample size was not sufficient to detect the effect of eczema on the risk of glioma. In our meta-analysis, we systematically evaluated existing data and provided more credible evidence for an association between eczema and glioma risk.The results from our study revealed an inverse association between eczema and risk of glioma . Finally, the studies were moderate to high quality according to the NOS.This is the first meta-analysis to demonstrate an association between eczema and risk of glioma. Our analysis had several advantages. First, we included 13 studies consisting of 10,897 glioma cases and 56,081 controls. Second, we included only case-control or cohort studies. Almost all of the cases in the included studies were histologically confirmed with well matched controls. Third, there was negligible heterogeneity  case–control or cohort study that investigated the association between eczema and glioma; (2) available patient data including the histological or clinical diagnosis ; (3) used OR or relative risk (RR) with CI to estimate the risk of glioma in individuals with a history of eczema compared to controls without the disease; and (4) recently published data in cases involving more than one published study of the same population. If studies had partly overlapping subjects, the study with the larger sample size was selected. The literature review and data extraction were independently performed by two investigators. Differences in opinion were resolved by discussion.Two authors independently extracted the data from each eligible study. The data included: (1) the name of the first author and affiliation; (2) publication year; (3) sample size; (4) nationality of the patients; (5) study design; (6) exposure assessment; (7) adjustment for potential confounders; (8) diagnostic criteria; and (9) reported outcome measures including OR or RR and 95% CI. Study quality was assessed by two investigators according to the NOS . This scP < 0.1 was considered significant [2I test, where 2I = 0 was indicative of absolute consistency and suggested low (25%–49%), moderate (50%–74%), and high (75%) threshold values [The meta-analysis was performed using the Stata 12.0 program. Cochran's Q-tests were performed to estimate heterogeneity among the individual studies. A nificant . The strd values . A fixedd values , 53. A P