Recently, a growing number of studies show that the killer cell immunoglobulin-like receptor (KIR) gene polymorphisms may play a role in the systemic lupus erythematosus (SLE) susceptibility. Nonetheless, the results were inconsistent. Thus, a meta-analysis was carried out by integrating multiple research to clarify the association between KIR polymorphisms and SLE susceptibility.The Web of Science, Embase (Ovid), PubMed, Elsevier Science Direct, the Chinese Biomedical Database and CNKI, Wanfang databases  were systematically searched to select studies on addressing the association between the KIR polymorphisms and susceptibility to SLE in populations. The odds ratio (OR) with 95% confidence interval (95% CI) was calculated.2DL1 =1.047, 95% CI=1.011–1.083) in all subjects. The KIR2DL3, KIR2DL5 were identified as protective factors for SLE in Asian populations , but not in Caucasians.A total of 10 published case-control studies involving 1450 SLE patients and 1758 controls were available for this meta-analysis. Results suggested that KIR2DL1 might be a risk factor for SLE (OR The meta-analysis results suggested that 2DL1 might be a potential risk factor and 2DL3, 2DL5 might be protective factors for SLE in Asians but not in Caucasians. It usually affects women more often than men in a ratio of 8:1 approximately. The peak age of SLE diagnosis ranges from 15 to 44 years old. Compared with other groups, women of child-bearing age are susceptible to the highest incidence of SLE. The etiology of SLE still remains unclear, but it is generally accepted that the onset of SLE is a multifactorial combined result, involving environmental, genetic, age, and hormonal factors. Aberrant innate immune responses play a significant role in the pathogenesis of SLE, contributing both to tissue injury via release of inflammatory cytokines as well as to aberrant activation of autoreactive T and B cells. Moreover, many studies have identified more than 30 robust genetic associations with SLE including killer cell immunoglobulin-like receptors (KIRs), IRF5, STAT4, PTPN22, TNFAIP3, BLK, BANK1, TNFSF4, and ITGAM.Systemic lupus erythematosus (SLE) is a systemic autoimmune disease with multiorgan inflammation. It is characterized by production of pathogenic autoantibodies and deposition of immune complexes in multiple organs, reflecting a global loss of self-tolerance, The KIR genes are clustered on the KIR locus located in the leukocyte receptor complex spanning a region of about 150 to 200 kb on chromosome 19q13.4. In humans, 17 highly homologous genes have been identified to date: 6 activating , 9 inhibitory , and 2 pseudogenes (KIR2DP1 and KIR3DP1). Recent studies have shown strong associations between KIR and human leukocyte antigen genes with susceptibility to autoimmune diseases such as rheumatoid arthritis,,10 multiple sclerosis, SLE,,13 type I diabetes (T1DM),,15 and type II diabetes (T2DM),,17 etc.The KIR receptors are members of the immunoglobulin family present on cell surface, expressed in NK cells21 and some T lymphocytes. To overcome the limitations on this issue, meta-analysis is an efficient method, which can integrate many previous small research studies and increase statistical power by pooling the results of single study. We performed this meta-analysis for the purpose of investigating the exact relationships between KIR polymorphisms and susceptibility to SLE.The association between KIR gene polymorphisms and SLE risk has been investigated by many case-control studies, but findings are not always consistent. Many of which were due to false-positive/negative studies, true variability in different populations, small sample sizes, low statistical power, or clinical heterogeneity are potential factors that could result in the discrepancies in these studies.22.1The meta-analysis was based on previous published studies, thus no ethical approval and patient consent are required.2.2 The following electronic databases were searched: the Web of Science, Embase (Ovid), PubMed, Elsevier Science Direct, the Chinese Biomedical Database (CBM) and CNKI, and Wanfang databases , using the following search terms:“systemic lupus erythematosus” or “SLE,” “The killer cell immunoglobulin-like receptors” or “KIR” or “KIRs,” “polymorphism” or “gene” or “genes” or “genotype” or “genotypes” or “genotyping” or “single nucleotide” or“allele.” All searched studies were retrieved, and their references were also scanned to identify additional relevant studies.This meta-analysis was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) guidelines.2.3Studies fulfilling the following inclusion criteria were qualified for this meta-analysis: case–control study; articles investigating the association between KIRs and SLE risk; sufficient published data available for estimating the odds ratio (OR) with 95% confidence interval (CI); and at least 2 comparison groups (SLE patient group vs control group) involved in a single study. The controls’ ethnic background and geographic area were comparable with the cases’. The publication was in English or Chinese. Exclusion criteria were as follows: research that did not meet the inclusion criteria; the study reported duplicated or useless data.2.4 This scale contains 9 items in total (1 point for each) in 3 parts: subject selection: 0 to 4; subject comparability: 0 to 2; exposure: 0 to 3. Study with NOS scores ≥7 was regarded as good quality .Newcastle–Ottawa scale (NOS) criteria was used to assess the methodological quality of all included studies.2.5Two investigators separately extracted the data with the standard protocol, any disagreement was addressed by discussion and resolved by consultation with the third investigator. The following items were extracted: first author, publication year, country of origin, ethnicity, genotyping method, KIR gene polymorphisms. The information is shown in Table 2.6P value <0.05 was deemed to be statistically significant. Heterogeneity was measured using the Q statistic, and P value < 0.1 was considered statistically significant. The extent of heterogeneity was further quantified using the I2 statistic, I2 values of 25%, 50% and 75% were defined as low, moderate, and high heterogeneity, respectively. When the P value was >0.10 or I2 value was <50%, the pooled OR was calculated using the fixed-effect model; otherwise, a random-effect model was used. The χ2 test was applied to examine whether the genotype distributions in the control group of each study were in Hardy–Weinberg equilibrium. Furthermore, we performed funnel plots and Egger test assessed potential publication bias, when P value <0.05 was considered statistically significant. Subgroup analyses were conducted by ethnic group. Sensitivity analysis was performed by excluding individual study to assess the stability of the results. All data analysis was performed with STATA 12.0 software.The strength of associations between KIR gene polymorphisms and SLE risk was assessed by OR with the corresponding 95% confidence intervals (CI), 33.1,22,23 were Caucasians, 6–28 were East Asians, 1 was South Americans. Owning to the insufficient sample populations available for the South American group, we conducted ethnicity-specific meta-analyses for Caucasian and Asian populations. Characteristics of the included studies are summarized in Table ,25,28,29 of 10 studies were of high quality (NOS score ≥7) and the other 4 studies were defined as moderate quality , which are shown in Table ,22–30The search strategy retrieved 126 potentially relevant studies . According to the inclusion criteria, 10 articles were included in this meta-analysis, which contained a total of 1450 SLE cases and 1758 healthy controls. A flow diagram of study selection process is illustrated in Fig. 3.22DL1 =1.047, 95% CI=1.011–1.083)  between studies was found in 2DL2, 2DL3, 2DL5, 2DS1–3, 3DS1, but heterogeneity in other KIR polymorphisms was not observed. Thus, the pooled OR was calculated by random effect model and fixed effect model, respectively.In our results, significant heterogeneity  and susceptibility to SLE in Caucasians. Meanwhile, there were 2 negative associations between 2DL3, 2DL5 , and SLE in Asians  . It is worth noting that in the sensitivity analysis of 2DL2, after omitting the Chen et al study, the result had substantially alteration and the findings should thus be treated with caution.In 2007, Pellett et al first reported that the frequency of KIR 2DS1 was significantly increased in SLE patients compared with controls. Therefore, we performed a subgroup analysis by ethnicity to identify the origin of heterogeneity. In the subgroup analyses, the results showed that 2DL1 might be a risk factor in Caucasians but not in Asians. Furthermore, the results also suggested 2DL3, 2DL5 might be protective factors in Asians but not in Caucasians, indicating that the association between KIR polymorphisms and the risk of SLE may be different in different ethnic populations. Various ethnic and environmental factors may influence susceptibility to disease, which is consistent with previous studies.,36 No evidence showed publication bias in this meta-analysis.Obvious heterogeneity between studies was detected, which may be attributed to different characteristics in different studies, quality of each research, different experimental designs and different genotyping methods, different geographic distributions with different genetic backgrounds. Second, we could not perform further subgroup analysis other than ethnicity, and even in the ethnic-specific subgroup analysis, we could only pool the results of Caucasians and Asians, respectively; south Africans were not possible due to limited data, and the division criteria of ethnic groups was crude, which may constrain the general application of our findings. Third, publication bias might occur for only including published articles, some relevant unpublished studies with null results were missed, Although we performed Egger regression test, we still could not remove the possibility of bias. Although our study has above limitations, this is the first meta-analysis focusing on the correlation between KIR gene polymorphisms and susceptibility to SLE. More importantly, we established a rigorous protocol and performed a statistical approach to integrate these inconsistent results through the whole process. Thus, the reliability of the results is guaranteed.Several limitations of our meta-analysis should be noted. First, we were unable to conduct further analysis to explore other risk factors due to the insufficient original data, such as environment–gene/gene–gene interaction, impact of different clinical subtypes or different course of SLE, and so on. Due to the above limitations, more research studies with larger sample sizes and more detailed information are needed to further validate and confirm our results.In conclusion, this meta-analysis shows 2DL1 might be a potential risk factor for SLE in Caucasians and 2DL3, 2DL5 might be protective factors for SLE in Asians. Thus, these genes may be applied to clinical early diagnosis of SLE as practical biomarkers.The authors thank all the anonymous reviewers and editors for their suggestions, which will be helpful for the authors to improve their paper.


P = 0.242), in the Caucasian population the G allele of mtDNA A10398G mutations may be a potential protective factor of PD . Further well-designed studies with larger samples are needed to validate these results.Many studies have researched the mitochondrial DNA (mtDNA) A10398G in Parkinson's disease (PD) to determine the association between mtDNA A10398G and PD, but the results of their research were not consistent. Therefore, we performed a meta-analysis to demonstrate the connection between mtDNA A10398G and the susceptibility of PD. We searched PubMed, Web of Science, Springer Link, EMBASE and EBSCO databases up to identify relevant studies. Through strict inclusion and exclusion criteria, at last, 9 studies  were included in our meta-analysis. We used the STATA 12.0 statistics software to calculate the pooled odds ratios (ORs) and 95% confidence intervals (CIs) to evaluate the genetic association between mtDNA A10398G and the risk of PD. We performed subgroup analysis to clarify the possible roles of the mtDNA A10398G polymorphism in the aetiology of PD in different ethnicities. Our meta-analysis indicates that although there was no significant association between mtDNA A10398G and PD in the Asian population (G vs. A: OR = 1.090, 95% CI = 0.939–1.284,  Parkinson's disease (PD), characterized by resting tremor, rigidity, slowness of movements and postural instability, is the second most common neurodegenerative disorder after Alzheimer's disease . AlthougHuman mitochondria have their own DNA, containing 37 genes in a single chromosome . Some gen = 451) were excluded. In addition, reviews or meta-analyses (n = 2) and animal studies (n = 28) were excluded, and 17 potentially relevant articles were considered worth further reviewing. Then, through screening the full text, studies with no sufficient allele data (n = 8) were excluded, and 9 potentially relevant articles were identified. Finally, after estimating the quality of the eligible studies, 9 articles  met our inclusion criteria and were included in our meta-analysis. Among them, 6 studies’ subjects were Caucasian, and 3 studies’ subjects were Asian. In these studies, they used PCR or TaqMan as their genotyping method and the genotype frequencies of the controls were in HWE. The detailed characteristics of the included studies are described in Table The selection flow diagram and exclusion reasons for articles are described in Figure yses n =  and anim2 = 69.8%, P = 0.001), so the random effect model was selected. We also performed subgroup analysis stratified by ethnicity . According to our results, although no significant association between mtDNA A10398G polymorphism and PD was found when polling all populations overall or in the Asian population , in the Caucasian population  the G allele of mtDNA A10398G mutations may be a potential protective factor of PD. The forest plot of mtDNA A10398G polymorphism in PD is shown in Figure The results of our meta-analysis are summarized in Table P = 0.602, Figure Begg's funnel plots and tests were used to estimate the publication bias. No publication bias was identified for the association between A10389G polymorphism and PD susceptibility . We all know that although the aetiological factor of PD is still obscure, the genetic factor has been well confirmed to have a great contribution to the development of PD. Many studies have suggested that the dysfunction of mitochondria could be involved in the development of PD –19. AddiP = 0.013). After stratification by gender, the increased risk appeared to be more significant in females . Of note, such a statistically significant association between A10398G polymorphism and the risk of PD was not always duplicated in other studies. D. Otaegui [P = 0.242), the G allele of mtDNA A10398G polymorphism seems to be a potential protective factor of PD in the Caucasian population .In this meta-analysis, we included a total of 9 eligible case-control studies above. Among the 9 included studies, six were in the Caucasian population and three were in the Asian population. In a study based on a regional Caucasian population from Spain, with a total of 450 cases and 200 controls, the 10398G allele was found to decrease the risk of PD with an OR (95% CI) of 0.52 (0.34–0.81)  detected Otaegui  reported Otaegui  observed Otaegui  case-con Otaegui  study (iMutation of mtDNA A10398G (A to G transition due to an amino acid substitution from Thr to Ala) was located in the ND3 gene that encodes one subunit constituting Complex I . ComplexSeveral reasons may help account for why the results between the Caucasian population and the Asian population are different. First, the A10398G polymorphism is differently distributed in the Caucasian population compared to the Asian population: the frequency of 10398G allele of mtDNA, which is found only 26% in Caucasian, is much higher in Asian population . The negWe are convinced that our meta-analysis has some special characteristics. Most importantly, compared with previous research, a larger sample size was used to estimate the effect, so our results are more reliable than the previous research results. In addition, according to a recent review on Parkinson's disease pharmacogenetics included all the meta-analysis published in PD , and A10There are also some limitations of our meta-analysis. First, as PD is a multifactorial disease, the potential interactions of genetic-genetic or genetic-environmental factors may influence the process of PD. Second, in view of the relatively small sample size and the limited study number, especially in the subgroup of the Asian population where only three studies were included, the power used to detect the real difference between cases and controls may not be very strong. Finally, as a result of insufficient data, other possible variables, including age of onset, gender or smoking, for the association between A10398G and PD could not be evaluated.In conclusion, our meta-analysis indicated that although the G allele of the mtDNA A10398G polymorphism has no significant relationship with PD in the Asian populations, it seems to be a potential protective factor of PD in the Caucasian population. Although a systematic investigation of the relationship between A10398G polymorphisms and the risk of PD could not be conducted because of the aforementioned limitations, it is important to gain a better understanding of the effect of A10398G polymorphisms on PD. Considering the limitations, investigations involving large case-control studies of different ethnicities should be conducted to clarify the possible roles of the mtDNA A10398G polymorphism in the aetiology of PD.Two investigators independently searched the related literature from PubMed, Web of Science, Springer Link, EMBASE and EBSCO databases published up to March 2017. We used the following key words:  and (“gene polymorphism” or “variant”) and (“Parkinson's disease” or “PD”). References of the eligible articles were also searched. Searching was restricted to human case-control studies.To be included, the studies must have met the following inclusion criteria: (1) case-control studies that studied the relationship between mitochondrial DNA A10398G and PD, (2) studies that included sufficient allele data for both patients and controls to calculate the odds ratio (OR) and 95% confidence interval (CI), (3) studies in which cases and controls were based on the same population, (4) studies that included participants diagnosed with PD according to the UK Parkinson's Disease Society Brain Bank criteria, and (5) full text studies written in English.The excluded criteria included the following: (1) case reports, editorials, meta-analysis or reviews; (2) studies with insufficient information about the mitochondrial DNA A10398G; (3) family-based studies; (4) studies whose subjects were ethnic minorities; and (5) studies in which the allele distribution in controls did not meet Hardy-Weinberg equilibrium (HWE).Two investigators independently selected the eligible articles and extracted the needed data. Inconsistencies between the two investigators were resolved by discussing with a third reviewer. The information extracted from each study included the first author's name, year of publication, ethnicity of the study population, and number of cases and controls for each allele.The Newcastle-Ottawa Scale (NOS)  was usedZ-test. P < 0.05 was considered to demonstrate a statistically significant difference. Heterogeneity between studies was examined by the Q-test  and I-square statistic tests (I2 < 50% shows little difference) [P-value of the Q-test > 0.05 or I2 < 50%, the fixed effects model was selected. Inversely, the random effects model was applied [P-value of Begg's test greater than 0.05 and articles with symmetrical distribution in the Begg's funnel plot were considered to have no significant publication bias [We used the Stata 12.0 software  to calculate all data analyses. In addition, we used the allele model to evaluate the association between mtDNA A10398G variant and PD risk. The OR and CI were used to assess the strength of the association between the variant and PD susceptibility. The pooled OR was measured using the ference) . The fix applied . We alsoion bias .


Supplemental Digital Content is available in the text Accumulating studies have explored the effect of thymidylate synthase enhancer region (TSER) variation on risk of pediatric acute lymphoblastic leukemia (ALL) with controversial results. Therefore, this quantitative meta-analysis was performed to assess synthetically the association of TSER variation with susceptibility to develop pediatric ALL.The PubMed, ScienceDirect, Google Scholar, Wanfang Database, and China National Knowledge Infrastructure were systematically retrieved to obtain the published case-control studies about the relationship between TSER variation and pediatric ALL risk. The quality assessment of the included studies was preformed and relevant information was collected. Odds ratios (ORs) and 95% confidence intervals (CIs) were applied to evaluate the strength of association.P = 0.41; 2R/2R vs 3R/3R: OR = 0.99, 95% CI = 0.84–1.16, P = 0.90; 2R2R vs 3R/3R+2R/3R: OR = 1.05, 95% CI = 0.92–1.21, P = 0.45; 2R/3R+2R/2R vs 3R/3R: OR = 0.97, 95% CI = 0.87–1.09, P = 0.63; 2R vs 3R: OR = 1.03, 95% CI = 0.92–1.15, P = 0.61). Similarly, there was no significant association existed in the stratification analyses according to ethnicity, control source, and quality score.This meta-analysis finally included 2681 children with ALL and 3854 matched controls from 11 investigations. The quantitative synthesis results found no significant association between TSER variation and susceptibility to pediatric ALL in overall comparisons under 5 genetic models (2R/3R vs 3R/3R: OR = 0.95, 95% CI = 0.84–1.07, This meta-analysis shows that TSER variation is not related to the development risk of pediatric ALL. Folate metabolism not only supplies the methyl group for proper DNA biosynthesis, it also provides the universal methyl donor for DNA methylation . Plenty of studies have clarified that low folate intake causes uracil misincorporation in the process of DNA replication reactions, resulting in DNA double-strand breakage, chromosomal deletion, and catastrophic DNA repair.,6 What is more, hypomethylation of DNA may also cause the activation of proto-oncogenes.,8 Emerging evidence has shown that variations in genes encoding folate-metabolizing enzymes disturb the balance of folate metabolism and have been associated with an altered susceptibility to cancer.–11Pediatric acute lymphoblastic leukemia (ALL) accounts for 30% of all malignancy diagnosed in children and 80% of pediatric leukemia.,13 Therefore, TYMS functions as an essential regulator in the process of DNA biosynthesis, repair, and methylation. The TYMS gene with 7 exons locates at 18p11.32. There are several functionally important variants in the TYMS untranslated regions, of which thymidylate synthase enhancer region (TSER) variation has been most widely investigated.–16 TSER, a tandem-repeat polymorphism, which includes double (2R) or triple (3R) repeats of a 28 bp sequence in the TYMS 5′-untranslated enhanced region, may be associated with an alteration in TYMS mRNA expression.,18 Considering the pivotal role of folate in the development of cancer and the potential influence of TSER polymorphism in the TYMS gene on DNA biosynthesis and methylation, it is reasonable that TSER variation might be related to susceptibility to develop malignancies. Increasing studies have found that TSER polymorphism has been linked to human various cancer risks, such as non-Hodgkin lymphoma, breast cancer, and colorectal cancer.–21 Recently, numberous investigations have explored the effect of TSER variation on development risk of pediatric ALL, yet the reported results remain controversial. The discrepancies among these studies may be ascribed to the genetic backgrounds difference and relatively small sample size in individual investigation. Therefore, a quantitative meta-analysis was performed to evaluate synthetically the association of TSER variation with pediatric ALL risk.Thymidylate synthase (TYMS) catalyzes the methylation of deoxyuridine monophosphate (dUMP) to deoxythymidine monophospate (dTMP), and maintains the balance of deoxynucleotide pool, which is needed for normal DNA replication and damage repair.22.1The PubMed, ScienceDirect, Google Scholar, Wanfang Databases, and China National Knowledge Infrastructure were systematically searched to screen reports about the association of TSER variation and risk of pediatric ALL utilizing the following keywords: “childhood” or “pediatric” or “children,” “leukemia” or “acute lymphoblastic leukemia” or “ALL,” “thymidylate synthase” or “TS” or “TYMS,” “polymorphism” or “mutation” or “variation” or “variant.” The latest literature search was performed on January 20, 2016 and there was no language restriction. In addition, the reference lists in the retrieved articles were screened to identify relevant investigations. Ethical approval was not necessary because this study was a meta-analysis.2.2The following inclusion criteria were applied for literature selection: case-control designed study; confirmed diagnosis for the case group of pediatric ALL; available genotypes distribution data for cases and controls. The letters, case reports, commentary, and review articles were excluded. If the same or overlapping data was reported by multiple articles, we chose the one with larger sample size.2.3 The NOS method, with a maximum score of nine points, includes 3 quality categories: selection, comparability, and exposure evaluation. Studies with more than 6 scores were identified as high quality. Any disagreement was resolved by reevaluation of the originally included studies.Two authors independently preformed the quality assessment of included studies according to the Newcastle–Ottawa Scale (NOS).2.4The information was collected from each eligible investigation independently by 2 authors: first author's name, publication year, country, ethnicity, sample size, control source, method used for genotyping, genotypes distribution data of the TSER variation in case and control group.2.5χ2 test was employed to check Hardy–Weinberg equilibrium (HWE) of genotypes distribution frequencies in control groups and P < 0.05 was considered as departure from equilibrium. The strength of association between TSER variation and pediatric ALL risk was measured by odds ratios (ORs) and 95% confidence intervals (CIs) under the homozygote model (2R/2R vs 3R/3R), heterozygote model (2R/3R vs 3R/3R), dominant model (2R/3R+2R/2R vs 3R/3R), recessive model (2R2R vs 3R/3R+2R/3R), and allele model (2R vs 3R), respectively. The χ2-test-based Q test was performed to estimate the heterogeneity between included studies. When P > 0.05, showing that no statistically significant heterogeneity existed, the fixed-effects model (Mantel–Haenszel) was employed to compute the pooled ORs; alternatively, the random-effects model (DerSimonian–Laird) was used. Stratification analyses were carried out based on ethnicity, control source, and NOS score. Sensitivity analysis was conducted by omission of studies deviated from HWE to assess the stability of combined results. Both qualitative funnel plot and quantitative Egger test were employed to assess publication bias. All the statistical tests were done with RevMan v5.3  and STATA v12.0 , and P < 0.05 was deemed to have statistical significance.The 33.1–33Table –29 4 studies on Asians,–33 and 2 investigations on mixed population.,24 Four investigations were hospital-based ,30,31,33 and 7 were population-based –29,32 designed when classified according to the control source. Four studies were divided into low quality with a NOS score of 4 or 5 points, and 7 with score 6 or greater were assigned as high quality. The alleles and genotypes distribution data of TSER variation in case group and control group are summarized in Table ,29Figure 3.2P = 0.41; 2R/2R vs 3R/3R: OR = 0.99, 95% CI = 0.84–1.16, P = 0.90; 2R2R vs 3R/3R+2R/3R: OR = 1.05, 95% CI = 0.92–1.21, P = 0.45; 2R/3R+2R/2R vs 3R/3R: OR = 0.97, 95% CI = 0.87–1.09, P = 0.63; 2R vs 3R: OR = 1.03, 95% CI = 0.92–1.15, P = 0.61). Similarly, no significant association was found in the stratification analyses according to ethnicity , NOS score , and control source   and 2R allelic carriers  in de Jonge et al study. To elucidate this inconsistency, a synthetical meta-analysis was conducted. In our study, no significant heterogeneity was observed among overall studies under all 5 genetic models. The combined data demonstrated that there was no significant association of TSER variation and risk of pediatric ALL in overall comparison under all genetic models. No significant association was found in the stratification analyses based on ethnicity, control source, and quality score. Our results were not in accordance with the conclusion published by Weng et al, which showed TSER variation might dedicate to significantly increased risk of childhood ALL . Since our study added several new investigations and included 2681 children with ALL and 3854 matched controls, which allowed for sufficient statistical power and more precise estimation, our conclusion is more reliable.To date, many epidemiological studies regarding the association of TSER variation with risk of pediatric ALL have been reported, but the published results remain controversial. Gast et alHowever, several limitations in our study need to be addressed in interpreting the results. First, due to data insufficiency, 2 relevant investigations were removed from the quantitative synthesis. Second, our analysis largely focused on single-factor estimates not adjusted for other confounders such as gender, lifestyles, and other potential factors, which may cause confounding bias and influence the combined results. The combined analyses of some subgroups may have no sufficient testing power to accurately assess the real association. In addition, the gene-environment interactions that may modify cancer susceptibility were not assessed in our study ascribed to the limited relevant information.5TYMS gene was not related to susceptibility to develop pediatric ALL. However, in the future, well-designed studies with more participants are demanded to verify this conclusion.In brief, this meta-analysis suggested that TSER polymorphism in


Evidence suggests that lifelong cumulative exposure to pesticides may generate lasting toxic effects on the central nervous system and contribute to the development of Alzheimer’s disease (AD). A number of reports indicate a potential association between long-term/low-dose pesticide exposure and AD, but the results are inconsistent. Therefore, we conducted a meta-analysis to clarify this association. Relevant studies were identified according to inclusion criteria. Summary odds ratios (ORs) were calculated using fixed-effects models. A total of seven studies were included in our meta-analysis. A positive association was observed between pesticide exposure and AD . The summary ORs with 95% CIs from the crude and adjusted effect size studies were 1.14  and 1.37 , respectively. The sensitivity analyses of the present meta-analysis did not substantially modify the association between pesticide exposure and AD. Subgroup analyses revealed that high-quality studies tended to show significant relationships. The present meta-analysis suggested a positive association between pesticide exposure and AD, confirming the hypothesis that pesticide exposure is a risk factor for AD. Further high-quality cohort and case-control studies are required to validate a causal relationship. Alzheimer’s disease (AD) is the most common progressive neurological disease and results in an irreversible loss of neurons3in-vitro models and animal studies suggests that long-term/low-dose pesticide exposure may lead to the neuronal loss in specific brain regions, resulting in subsequent cognitive impairment, decreased memory and attention, and loss of motor function1012The aetiology of late-onset AD remains largely unknown but is believed to be multifaceted, resulting from both genetic and environmental factors7Pesticides are known neurotoxins. Although the mechanisms underlying the influence of pesticides on neurodegenerative disease remain to be elucidated, the role of long-term/low-dose exposure to pesticides such as paraquat, dieldrin, organochlorine and organophosphates has long been suspected. Most pesticides share a number of features, such as the ability to induce oxidative stress, mitochondrial dysfunction, α-synuclein fibrillization and neuronal loss15In order to investigate the epidemiological relationship between AD and pesticide exposure, we performed a comprehensive systematic review and meta-analysis of published cohort and case-control studies on the association between pesticide exposure and development of AD.Our initial search of all databases retrieved 1,526 studies; however many of these were duplicates. After review of titles and abstracts, we identified 104 pertinent articles. We excluded reviews, editorials, nonhuman studies, and case reports; we also excluded publications that explored the association between serum pesticide levels and risk of AD, as these data could not be meta-analysed because of their different formats in the reporting of results. After excluding those studies that failed to meet the inclusion criteria, a total of seven publications were selected, providing data on three cohort1217192021The seven eligible studies included 6,835 participants, of which 1,050 were AD patients. The main characteristics of the selected studies are presented in P = 0.88 and 2I = 0.0%), indicating that the results across the selected articles were statistically homogeneous. We used a fixed-effects model to calculate summary odds ratios (ORs) and 95% confidence intervals (CIs). The summary OR of pesticide exposure and AD was 1.34 , suggesting a significantly positive association, the forest plot was shown in Heterogeneity between the included studies did not exceed that expected by chance , but not in case-control studies . When stratified by method used to assess exposure, the OR of the self-reported group was 1.37 , and the OR of the proxy-reported group was 1.24 .When stratified by control source, no significant association was observed for studies utilizing hospital controls . In contrast, we observed a significant association between pesticide exposure and AD in studies utilizing community and population based controls . When stratified by study quality (score < 7 or ≥7), significant association was observed for study scores ≥7 , though the association observed for study scores < 7 was not significant . Moreover, in order to avoid the residual confounding variable of age, we also conducted subgroup analyses based on whether or not the pooled ORs had been adjusted for age. The results revealed an increased risk in the age-adjusted group , but not in the non-age adjusted group . In addition, pesticide exposure was associated with AD for studies in which the mean age of patients with AD was less than 80 years old , while non-association was noted among studies in which the mean age of patients with AD was greater than 80 years old . Furthermore, pesticide exposure was found to be associated with AD for studies whose results had been adjusted for more than three variables. The results of subgroup analyses are shown in Sensitivity analyses were conducted to evaluate the sensitivity of our conclusions. First, the random-effect model and fixed-effect model were compared with the quality-effect model, and the conclusions remained unchanged. Second, we omitted each study from the analysis one-by-one . The conclusion was not drastically changed upon this analysis, and ORs ranged from 1.26  to1.38 . All results were either significant or of marginal significance. The results of the leave-one-out analysis are shown in P = 0.66; Begg, P = 0.76), for studies reporting crude effect size , or for studies reporting adjusted effect size . Thus, no significant evidence of substantial publication bias was observed in this study.Neither Egger’s test nor Begg’s test suggested any evidence of publication bias for pesticide exposure  significantly increases levels of amyloid- β precursor protein (AβPP) and β-site AβPP-cleaving enzyme 1(BACE1), impairing the clearance and extracellular degradation of amyloid-β peptides28in vivo study revealed that some pesticides may disrupt metabolic pathways involved in the homeostasis of amyloid-β, causing a significant increase in amyloid-β levels in the cortex and hippocampus, as well as increased memory loss and reduced motor activity in experimental animals12The relationship between pesticide exposure and risk of AD has attracted an increasing amount of attention in recent years. Pesticides are well-known neurotoxins and are associated with many neurodegenerative disorders, including mild cognitive impairment and dementia, which are strongly linked to AD. Mild cognitive impairment is a prodromal phase of cognitive decline that may precede the emergence of AD. Some research has suggested that mild cognitive impairment and late-onset AD are essentially part of the same pathophysiological process, sharing a number of etiological factors26P = 0.88 and 2I = 0.0%), indicating that the selected articles were statistically homogeneous, and that the results exhibited relative reliability. Sensitivity analyses yielded similar results, indicating that the relationships were relatively stable. Moreover, our results were consistent with an ecological study25To our knowledge, the present study is the first comprehensive meta-analysis combining data from cohort and case-control studies to investigate the possible relationship between pesticides and AD. The results of our analysis suggest a positive association between overall pesticide exposure and AD , without heterogeneity  due to the mental decline of patients with AD, and this type of information may increase the degree of misclassification bias. Such retrospective data are less reliable, and recall bias becomes unavoidable. Furthermore, the exposure information of both patients control and participants was provided by surrogate respondents in case-control studies; thus, the presence of non-differential bias may result in an underestimation of the effect sizeIn epidemiological studies, adjustments should be conducted to reduce potential confounding variables and achieve more reliable conclusions. Our present study showed that the summary OR of studies reporting adjusted effect sizes was 1.37 , exhibiting a higher risk than those reporting crude effect sizes . In our meta-analysis, age was an important factor associated with the risk of AD and pesticide exposure duration32Potential bias is the major challenge for meta-analyses of observational studiesThe method of AD diagnosis is less likely to be a source of bias in the present study because our included studies used only one criterionThe highlight of our study is that it is the first comprehensive meta-analysis to investigate the possible relationship between pesticide exposure and the risk of AD. A second advantage is that no heterogeneity was observed among the original studies, suggesting that the result of the present study reveals a reliable relationship. A third advantage is that the sensitivity analysis of our present study did not substantially modify the association between pesticide exposure and AD, indicating stable results. Moreover, our conclusions are consistent with most previous studies, including internal pesticide exposure studies and those regarding the influence of pesticides on dementia and mild cognitive impairment.However, several limitations exist in the present meta-analysis. First, as we did not search for unpublished studies or original data, publication bias may be inevitable, even though no significant evidence of publication bias was observed. Second, exposure bias was unavoidable because the methods of exposure determination in the original studies ranged from self-administered questionnaires to proxy reports. Third, due to lack of relevant studies, the relationship among AD and duration of pesticide exposure, and patient occupation was not thoroughly investigated. Lastly, we did not thoroughly investigate potential differences with regard to exposure to specific compounds or functional group of pesticide exposure in our subgroup analyses.The results of our meta-analysis suggest a significant positive association between pesticide exposure and incidence of AD. These findings provide powerful evidence supporting the hypothesis that pesticide exposure is related to an increased risk of AD. Further prospective cohort studies and high-quality case-control studies with improved methods for estimating cumulative pesticide exposure and documenting cases of AD are required to validate the existence of a causal relationship.We searched PubMed, EMBASE, and Web of Science databases for all English-language cohort and case-control studies published in peer-reviewed journals through April, 2016. PubMed search terms were (“Alzheimer Disease” OR “Dementia” OR “Alzheimer Disease”[Mesh]) AND  AND “Risk”. Similar search terms were used for EMBASE and Web of Science. References from eligible articles were also examined for additional studies. We conducted our meta-analysis according to the PRISMA checklists and followed the relevant guidelinesWe selected studies on the basis of the following criteria: cohort or case-control study design, the exposure of interest being pesticides and the outcome being the incidence of AD, and articles that reported at least one effect size with 95% CI relating pesticide exposure to AD or enough data to calculate effect size. Serum pesticide exposure data were excluded, as these data could not be meta-analysed because the results were reported in different formats. Studies that utilized mortality data for ascertainment of AD were also excluded, as AD may be frequently underreported on death certificates.Two investigators (D.Y. and Y.Z.) extracted the data and evaluated the eligibility of potential studies according to the guidelines for meta-analysisThe Newcastle-Ottawa ScaleQ and 2I statisticsP value for the Cochran Q statistic indicates a significant level of heterogeneity. The 2I metric describes the percentage of total variation among studies that is due to heterogeneity rather than chance342I values were considered to be 25%, 50%, and 75%We chose ORs as a common measure to assess the relationship between pesticides exposure and AD because when the outcome is rare, ORs, hazard ratios (HRs), and relative risks (RRs) provide similar estimates of riskTo explore the potential effects of specific study characteristics on the association between pesticide exposure and AD, subgroup analyses were conducted according to source of controls, study quality, methods of exposure assessment, age, and adjustments. Publication bias was evaluated by Egger’s test and Begg’s test43P values) were two-sided, and P < 0.05 was considered statistically significant.Except for quality-effect modelling (conducted with MetaXL version 2.2 software), all statistical analyses were conducted with Stata 12.0 software (StataCorp). All reported probabilities .


CD44v6 has recently been reported as a biomarker for colorectal cancer. However, the clinical and prognostic significance of CD44v6 in colorectal cancer remains controversial. Therefore, we performed a meta-analysis to clarify this issue. A comprehensive literature search was performed using Medline, Embase and Web of Science, and the statistical analysis was conducted using Stata software. A total of twenty-one studies including 3918 colorectal cancer cases were included. The pooled analysis showed that CD44v6 overexpression in colorectal cancer was an independent prognostic marker correlating with lower 5-year overall survival rate . CD44v6 overexpression was also associated with more lymph node invasion , and advanced Dukes stage . In addition, while excluding Zolbec's study, CD44v6 overexpression was associated with distance metastasis . Taken together, this meta-analysis suggested that CD44v6 is an efficient prognostic factor in colorectal cancer. Although the treatments for colorectal cancer (CRC) have developed rapidly in recent years, CRC still remains one of the most commonly diagnosed cancers and the second leading cause of cancer-related death worldwide , 2. CRC As a compelling stem cell marker, CD44 has been reported to play important roles in tumor initiation and metastasis . CD44 isAmong the isoforms of CD44, CD44v6 has evoked special interest. CD44v6, like all other isoforms, contains a hyaluronan-binding site in its extracellular domain and serves as a major cell surface receptor for hyaluronan . TherefoMany studies have been conducted to explore the relationship between CD44v6 and the prognosis of CRC –32. MostThree hundred and seventy potential papers were identified initially using the search strategy above, 326 of which were excluded after reading the titles and abstracts. After reading full texts, we excluded another 23 studies due to no usable data and not-immunohistochemistry method. Finally, 21 studies met the inclusion criteria and included in this meta-analysis, including 3918 cases. The detailed literature selection procedure was described in Figure The association between CD44v6 and several clinicopathological features of CRC was illustrated in Table A total of thirteen studies reported the association of CD44v6 and 5-year OS rate of CRC, while only three studies reported the correlation of CD44v6 with 5-year DSF rate of CRC. The pooled analysis revealed that CD44v6 overexpression was significantly associated with poor 5-year OS rate in a fixed-effects model  Figure , 5-year Influence analysis was performed to assess the influence of each individual trial on the pooled ORs regarding of 5-year OS by sequential omission of individual trial. The analysis suggested that no individual trial could statistically significant affected the pooled ORs of 5-year OS Figure . HoweverPotential publication bias was examined by funnel plots in the analysis of CD44v6 expression and 5-year OS rate. The shapes of the funnel plots was symmetric, suggesting no obvious publication bias. Furthermore, the p values assessed by the Begg's test and Egger's test were all greater than 0.05, indicating no obvious publication bias among these studies regarding the OR for 5-year OS rate  are regarded to be significantly responsible for initiation, growth, metastasis and recurrence of various tumors . Many ceTo our knowledge, this is the first meta-analysis which systematically evaluated the association between CD44v6 expression and the risk of CRC and its clinicopathological parameters. Our meta-analysis included 21 studies with a total of 3918 cases, reflecting a large sample size and strong statistical power. In this meta-analysis, we found that overexpression of CD44v6 is related to lymph node metastasis, Dukes stage of CRC, and distance metastasis. Furthermore, CD44v6 overexpressed CRC patients was of lower 5-year overall survival in comparison with negative cases. In conclusion, overexpression of CD44v6 and its clinicalpathological features are closely related in CRC. Our results provide further support to the potential translational significance of CD44v6 in CRC. Todaro and colleagues  found thAlthough this meta-analysis aimed to provide the best possible estimate of the correlation between the clinical significance of CD44v6 in CRC, it may have limitations. First, the number of included articles was relatively small with only about 3918 cases. Second, there was between-study heterogeneity in most of the analysis, this may be due to the different concentrations of antibodies, the different cut-off values of CD44v6 expression used in the studies. Third, although immunohistochemistry was the most commonly applied method for detecting CD44v6 in situ, RT-PCR method had also been used for the evaluation of the levels of CD44v6 mRNA expression in CRC tissue. However, these studies measuring CD133 mRNA level by RT-PCR were not yet included in this meta-analysis. Finally, publication bias may have occurred though the funnel plot did not show it since negative findings were likely to be unreported.In summary, despite the limitations listed above, this meta-analysis shows a significant correlation between CD44v6 overexpression and 5-year OS rate as well as metastasis in CRC patients. CD44v6 may have prognostic significance for CRC patients based on currently obtained data. Further larger well-designed prospective studies are warranted to confirm these findings from our study.Eligible studies were retrieved by searching the following database: Medline, Embase, and Web of Science. The search strategy included the following keywords: “CD44V6”, “CD44 variation 6”, “colorectal cancer”, “colon cancer”, “rectal cancer”. In addition, the reference list of each primary study and of previous systematic reviews were also manually searched to avoid missing studies.All eligible studies evaluated the association between CD44v6 expression and the risk of CRC were selected in this meta-analysis. Studies meeting the following inclusion criteria were included: 1) CD44v6 expression was detected in CRC by immunohistochemistry; 2) studies about the relationship between CD44v6 expression and the clinicopathological features or prognosis of CRC; 3) studies regarding the prognosis of CRC provided sufficient data to estimate hazard ratios (HRs) for overall survival (OS) or disease free survival (DFS). Those meta-analysis, reviews or systematic reviews, abstracts, letters and those provided not sufficient data and not using immunohistochemistry were exluded. We did not assess the methodologic quality of the included studies, given that quality scorings of observational studies in meta-analysis is controversial.http://getdata-graph-digitizer.com).All data were independently extracted by two investigators (JW and WS), and disagreements in data extraction were resolved by discussion. The following data were recorded from each article: the name of first author, publication year, country of the population studied, number of cases, duration of follow-up, T category, N category, distant metastasis, tumor size, histology, tumor grade, TNM stage, Dukes stage, and most importantly the 5-year OS rate and 5-year DFS rate. For those studies which didn't provide 5-year OS and DFS directly, Kaplan–Meier curves were read by GetData Graph Digitizer . Comparisons of CD44v6 expression in groups with different clinicopathologic features and prognosis of CRC were assessed by pooled estimates of risk ratio as well as the 95% CI. p values less than 0.05 were considered statistically significant. The chi-squared (χ2) test and I2 statistics was calculated to assess between-study heterogeneity. Fixed-effects models (Mantel-Haenszel) were used when there was no between-study heterogeneity, otherwise, the random effect models (DerSimonian and Laird) were used. A funnel plot was used for estimating the potential publication bias, while the degree of asymmetry was determined by Begg's and Egger's test. Influence analysis was conducted by omitting each study to find potential outliers. Two authors (JW and WS) performed the statistical analysis independently and got the same results.


Background. Studies revealed that metabolic factors might contribute substantially to osteoarthritis (OA) pathogenesis. There has been an increasing interest to understand the relationship between knee OA and the metabolic syndrome (MetS). The purpose of this study was to explore the association between metabolic syndrome and knee osteoarthritis using meta-analysis. Methods. Databases, including PUBMED, EMBASE, and the Cochrane Library, were searched to get relevant studies. Data were extracted separately by two authors and pooled odds ratio (OR) with 95% confidence interval (CI) was calculated. Results. The meta-analysis was finished with 8 studies with a total of 3202 cases and 20968 controls finally retrieved from the database search. The crude pooled OR is 2.24 (95% CI = 1.38–3.64). Although there was significant heterogeneity among these studies, which was largely accounted for by a single study, the increase in risk was still significant after exclusion of that study. The pooled adjusted OR remained significant with pooled adjusted OR 1.05 . No publication bias was found in the present meta-analysis. Conclusions. The synthesis of available evidence supports that metabolic syndrome increases the risk for knee osteoarthritis, even after adjustment for many risk factors. Knee osteoarthritis (KOA) is a prevalent chronic joint disease affecting about 28% of the population in the USA at 45 years of age. It is estimated to rise up to 37% in adults aged over 65 years  within tRecently, studies revealed that metabolic factors might contribute substantially to OA pathogenesis . There h"Osteoarthritis"[Majr]) AND "Metabolic Syndrome X"[Majr], “osteoarthritis metabolism syndrome”, and “osteoarthritis metabolic syndrome”. No language restrictions were applied. A reference list of relevant papers was also screened.The authors did an intensive search on PUBMED, EMBASE, and the Cochrane Library for published studies until June 3, 2015. Relevant available articles were obtained using key words such as (A study was included if it met the following inclusion criteria: (1) It is cohort, case-control, or cross-sectional study investigating the association between MetS and risk of OA; (2) relative risk (RR) or odds ratios (ORs) and the corresponding 95% confidence intervals (CIs) could be directly extracted or calculated from the available data; and (3) if there is duplicated data, the larger sample size was adopted. Two investigators independently applied the selection criteria to each reference identified by the search strategy. Any discrepancies regarding eligibility or quality were resolved by the third reviewer.The following information was collected from each available article: year of publication, ethnicity of the studied population, numbers of cases and controls, mean age of case and control groups, and MetS criteria and OA criteria adopted adjusted OR and adjusted variables. All articles were independently reviewed by two investigators. Data was extracted independently and entered into separate databases. Results were compared and any discrepancy was resolved by the third investigator as well.The National Cholesterol Education Program-ATPIII (NCEP-ATPIII) definition was used to define metabolic syndrome (MetS) . MetS isKellgren and Lawrence grade (K-L grade) classified KOA into four grades . Radiographic knee OA is defined as K-L grade 2 or above . The AmeI2 >50%, heterogeneity was significant, and random-effects model was applied to estimate the pooled ORs and 95% CI. We considered I2 of <50% as “heterogeneity might not be important,” and a fixed-effects model will be conducted. Cochrane Collaboration's Review Manager Software Package  was used for the meta-analysis.The crude pooled OR was calculated. Adjusted ORs controlling for potential confounding factors in the greatest degree were also extracted whenever available. The adjusted OR was converted by using the natural logarithm and the SEs and their corresponding 95% CIs were calculated from these logarithmic numbers. Pooled adjusted OR was also calculated. When A flow chart of literature search and study selection is shown in p = 0.555) . The Galbraith plot showed that the study by Puenpatom and Victor largely accounted for heterogeneity , with the OR attenuated but remaining statistically significant .The pooled ORs result suggested that MetS seemed to significantly increase overall KOA risk  . The Egg= 0.555) . Howeverogeneity . After ep < 0.00001)  and there was no publication bias (p = 0.649)  . No sign= 0.649) .KOA is the most prevalent form of arthritis and a major cause of pain and disability that affected 151 million individuals globally in 2000 , 5. MetSThis study showed that MetS increases the risk of KOA, even after adjusting many of the risk factors. In our meta-analysis, one article was excluded for not providing the data for analysis. However, in the excluded article, it was indicated that MetS was prevalent in 59% of patients with OA, whereas it was 23% in the population without OA . Thus, eOA and MetS share the same mechanisms of inflammation, oxidative stress, common metabolites, and endothelial dysfunction , 19–21. κB signalling pathway which plays a critical role during OA development [As parts of the metabolic syndrome, insulin resistance, hyperglycaemia, and hyperinsulinemia are also strongly related to OA pathogenesis. In clinic, diabetes has been proved to be an independent predictor for osteoarthritis by the population-based cohort study followed over 20 years . There ielopment , 33.All of the above highlight the association between MetS and KOA. MetS not only increases the susceptibility to KOA but also exerts influence on its progression and prognosis. Yoshimura et al. demonstrated not only a dose-response relationship between MetS and KOA but also the relationship between MetS components accumulation and KOA progression and occurrence , 34. OurThis study has several limitations. First, the number of studies evaluated was very small. Future studies should include a more comprehensive analysis of the topic. Second, although genetic risk factors for OA have been found to contribute to the establishment and progression of this condition , in thisFor the first time, with meta-analysis, the synthesis of available evidence supports that metabolic syndrome increases the risk for knee osteoarthritis, even after adjustment for many risk factors. As a result, OA is a heterogeneous disease and metabolic factors contribute substantially to its pathogenesis. Furthermore, understanding that MetS contributes to KOA patients will be useful in assessing KOA patients' individual conditions and selection of precision therapeutic strategies.


To clarify the relationship between serum, dietary, and urinary potassium and the risk of type 2 diabetes mellitus (T2DM).We searched PubMed and EMBASE through January 6, 2017 for studies reporting risk estimates on the association of potassium measurements and the risk of T2DM. The summary risk estimates were obtained through a random-effects model. Dose-response analysis was conducted.non-linearity > 0.05).Eight studies involving 5,053 cases and 119,993 individuals were included. A trend toward significance was found in the highest versus lowest meta-analysis on serum potassium and T2DM risk ; moreover, the RR per 1 mmol/L increase in serum potassium was 0.83 (95% CI 0.73–0.95). A non-significant association of dietary potassium and T2DM risk was detected . A similar non-significant association was found for urinary potassium and T2DM risk . Evidence of a linear association between serum, dietary, and urinary potassium and the risk of T2DM was found (all PLow serum potassium increases the risk of T2DM in a linear dose-response manner; nevertheless, neither dietary potassium nor urinary potassium shows any association with the risk of T2DM. However, these findings should be interpreted with caution due to limited studies. According to the pooled data of the International Diabetes Federation from 219 countries and territories, diagnosed diabetes reached 381.8 million cases in 2013 and is projected to reach 591.9 million by 2035 . In addiPotassium is a chief cation in the intracellular fluid and plays a critical role in water- electrolyte balance. In the past decade, some large scale epidemiological studies have explored the association between serum, dietary, and urinary potassium and the risk of T2DM –14. HoweTo our knowledge, a systematic review and dose-response meta-analysis of potassium measurements and T2DM risk is not currently available. Therefore, we conducted this study to clarify the relationship between serum, dietary, and urinary potassium and the risk of T2DM.As shown in the flow chart Figure , 7,480 cThe characteristics of the eight identified studies are summarized in Table P > 0.05; When it came to quality assessment, six studies had 7 or 9 stars, whereas the remaining two had 5 stars, indicating the quality of included studies is generally good . In addiPheterogeneity < 0.01, I2 = 76.7%, 5 studies)  was found by ignoring two studies [2 = 18.59, Pgoodness-of-fit = 0.07, 5 studies). We found evidence of a linear trend (Pnon-linearity = 0.10) by using restricted cubic spline model [Five studies , 13, 14  studies  with muc studies , 13, 14  studies  the mean2 = 0.0%, Pheterogeneity = 0.52, 6 studies) . Evidence of a linear association was found      was founstudies) .Our dose-response meta-analysis demonstrated that serum potassium levels were linearly associated with the risk of T2DM, with each 1 mmol/L increase in serum potassium lowering the risk by approximately 17%. Despite the results of the highest versus lowest meta-analysis on serum potassium and the risk of T2DM being statistically non-significant, we observed a trend toward significance. Moreover, a statistically significant inverse association was identified after excluding the two outliers. Heterogeneity appeared to be related to a difference in the average age of subjects. Our study did not detect a significant relationship between dietary or urinary potassium and the risk of T2DM.Our study is the first meta-analysis to confirm this association quantitatively and is consistent with the results of two previous qualitative reviews , 20 whicP < 0.01). These results are supported by another randomized clinical trial of 3790 non-diabetic participants [The effects of thiazide diuretics on the risk of T2DM have been discussed for over half a century. A positive association with increased blood glucose , 22 as wicipants , where ep < 0.05). Further experimental evidence indicates glucose intolerance induced by hypokalemia could be reversed by potassium supply [A large number of experimental studies have been conducted to determine whether experimentally induced hypokalemia is directly associated with glucose metabolism. In a study performed by Rowe et al , seven hm supply –31. Morem supply  by increm supply . NeverthSeveral animal model studies –36 suppoIn addition, further studies have concentrated on a molecular mechanism. A meta-analysis of 42, 573 individuals in East Asia , showed The significant linear inverse relationship between serum potassium and T2DM risk may be explainable by the above-mentioned mechanisms. For dietary and urinary potassium, the null associations are probably based on the different measurements of potassium intake. Though food frequency questionnaires (FFQ) and 24h urinary potassium levels are the two most accurate measurements of potassium intake , these mOur study has several prominent strengths. Firstly, all included studies are prospective cohorts which result in fewer recall and selection biases. Moreover, we employed the most fully adjusted models to decrease the potential confounding factors. More information could be obtained from the combined use of categorical and dose–response statistical methods.There are also limitations. First, the number of included studies was limited. Thus, we did not perform subgroup analysis to explore the potential effect modifiers. In addition, several studies have the same first or prominent co-author, the validity of our conclusions may thus be influenced due to the clustering around one author. Second, owing to the nature of observational studies, the residual confounding factors are not fully excluded. Furthermore, dietary potassium was assessed by self-reported FFQs, which might have led to measurement errors. Considering the inconsistent diagnostic criteria of T2DM across studies, both under-reporting and over-reporting are possible. Finally, except for one study performed in Japan , other sIn conclusion, our meta-analysis suggests that low serum potassium increases the risk of T2DM in a linear dose-response manner, especially in relatively young individuals. However, neither dietary potassium nor urinary potassium shows any association with the risk of T2DM.Our results cannot prove causality due to the nature of epidemiologic studies. Future clinical trials are needed to determine if addressing low serum potassium decreases T2DM risk.Our meta-analysis was conducted following the MOOSE  guidelines . We searCitations were independently reviewed by two investigators (Y.P. and A.W.). We included prospective cohort studies on the association between potassium measurements  and the risk of T2DM in the general population that reported the adjusted risk estimates  and corresponding 95% confidence intervals (CIs). The most recent article was considered for inclusion when multiple articles derived from the same cohort. We excluded the study of gestational diabetes, new-onset diabetes after transplantation and type 1 diabetes. A consensus was reached by discussion in case of any discrepancy.A standard extraction form was used in the process of data extraction by two independent investigators (K.L. and Q. M.). Individual study was extracted with following information: first author, publication year, study location, age range, mean age, gender, sample size, incident T2DM cases, follow-up duration, categories of potassium, diagnostic criteria of T2DM, adjusted risk estimates and corresponding 95% CIs and adjustment variables.We adopted the Newcastle–Ottawa Scale, which is frequently selected for observational studies in meta-analyses, to evaluate the methodological quality of the selected studies. The total score ranges from 0 to 9 based on selection of participants and exposure, comparability on confounders, evaluation of outcomes, and sufficient follow-up. The total score exceeding six is regarded to be high-quality.−1or mg kJ−1 day−1as the unit of dietary potassium, we unified the unit to mg day−1 by multiplying average amount of daily energy intake provided in the study. The total 24h urinary potassium was approximately estimated by multiplying the spot urinary potassium (mmol/L) with the mean urinary volume per day if it was not available in the study [2 statistic  were calculated for evaluating heterogeneity across studies.We calculated summary risk estimates from each individual study using a random-effects model. Hazard ratio, odds ratio and risk ratio were directly regarded as equivalents. For studies , 14 withhe study . For sevhe study –14, 17 whe study . Both thhe study   were used for comparing the highest versus lowest category of dietary potassium.A random-effects dose-response meta-regression model described by Orsini et al  was carr studies , 16 wereP < 0.05 under a two-sided test.To ensure the stability of the results and to explore the possible sources of heterogeneity, sensitivity analyses were performed by omitting one study in turn. Due to the limited studies, subgroup analysis was not conducted. Begg’s and Egger’s tests were adopted to assess publication bias. Funnel plots were also mapped . We carr


Effect estimates were higher for studies with direct BZ/BNA exposure  , and studies that identified BZ/BNA-associated bladder cancer with SMR/SIR ≥4.7 . Effect estimates were similar for studies with and without concomitant occupational exposure to chromium, asbestos, arsenic, or bis(chloromethyl) ether. The cumulative meta-analysis showed that the evidence of association between occupational BZ/BNA exposure and lung cancer has been stable since 1995. Although the results of this meta-analysis have the potential for confounding by smoking and heterogeneity, our findings suggest that a finding of lung cancer following occupational BZ/BNA exposure should be considered to be a potential occupational disease.Benzidine (BZ) and beta-naphthylamine (BNA) have been classified as definite human carcinogens for bladder cancer by the International Agency for Research on Cancer. However, the epidemiological evidence for an association between exposure to BZ and/or BNA and lung cancer has been inconclusive. We conducted a systematic review and meta-analysis to determine the risk for lung cancer among workers exposed to BZ/BNA. A systematic literature search was conducted to identify studies that had reported occupational BZ/BNA exposure and the outcome of interest (lung cancer death and/or incidence). Meta-analyses were performed using random effects models to combine standardized mortality ratios (SMRs) or standardized incidence ratios (SIRs). We identified 23 retrospective cohort studies including 1745 cases of lung cancer; only one study reported smoking-adjusted lung cancer risk. A significantly increased lung cancer risk  was observed by combining all studies, with significant heterogeneity among studies (I This is due to inconsistencies in tumor site studies.11In addition to the carcinogenicity of BZ and BNA on the bladder, some occupational epidemiological studies have observed carcinogenic risks at sites besides the bladder  statement.16 This systematic review protocol was registered with the International Prospective Register of Systematic Reviews (PROSPERO) database (registration number: CRD42014010250) and published in an academic journal.17 This systematic review and meta-analysis is based on a method of prospective systematic reviews.18This systematic review and meta-analysis was conducted following the Meta-analysis of Observational Studies in Epidemiology (MOOSE) guidelinesWe included subjects with unequivocal evidence of occupational exposure to BZ and/or BNA, such as dyestuff workers, workers from BZ/BNA manufacturing plants, leather tannery workers who used BZ-based dyes, and rubber industry workers exposed to BZ/BNA present as a contaminant in antioxidants used in manufacturing. Subjects who worked at the same factories but were not exposed to either BZ or BNA were excluded.Eligible studies were comparative observational studies that reported occupational BZ/BNA exposure and the outcome of interest (lung cancer death and/or lung cancer incidence). Retrospective cohort studies , prospective cohort studies, and case-control studies were included in this review.For cohort studies, ascertainment of exposure to BZ and/or BNA was based on written records of exposure measurements or work history. For case-control studies, BZ and/or BNA exposure was ascertained by secure records , structured interviews with blinding to case/control status, interviews without blinding to case/control status, or written self-reports.For cohort studies, use of a comparator was not a requirement for inclusion. For case-control studies, the control group must have included subjects who had no history of lung cancer.Our outcomes were lung cancer death and/or lung cancer incidence based on clinically confirmed diagnoses . Effect measures included standardized mortality ratio (SMR), standardized incidence ratio (SIR), and odds ratio (OR) for the association between BZ/BNA exposure and lung cancer risk. SMRs and SIRs were based on an external comparison group , and ORs were based on a population- or hospital-based control group.The search strategies were carried out by the research team and an expert librarian (TS). No language restriction was enforced, conditional to the provision of an English abstract. A date restriction was not imposed. A comprehensive search of databases from each database’s earliest inclusive dates to September 11, 2014, was conducted; a subsequent update search  was also conducted. The databases included MEDLINE, Excerpta Medica DataBase (EMBASE), and Cumulative Index to Nursing and Allied Health Literature (CINAHL). Search terms included controlled vocabulary and text-words, and details of the search strategy for MEDLINE are provided in 19 For studies reporting data on incidence and mortality, incidence data were selected.We assessed for inclusion all titles and abstracts identified during the literature search. Two of the authors (KT and KS) independently examined the search results for potentially eligible studies. Disagreements were resolved through consultation with a third author (KO). For studies that appeared to meet the criteria for this review, we obtained and examined the full-text articles. If multiple reports of the same study were encountered, they were used only once; the record containing the most data  was used. This usually meant using the most recently published reference, but if lung cancer risk data was only noted an earlier paper, that information was used.20 If a study used both national and regional populations to compute the expected deaths/cases, the regional results were used, because a regional population of study subjects is preferable to the national population for controlling geographic differences in disease incidence.21 Data on the following study characteristics were also extracted where available: year of publication, proportion of males, duration of follow-up, duration of employment, country and geographic area, industry type, occupational exposure to chemicals (including BZ and BNA), information on cigarette smoking, and information relating to quality assessment. Also, information concerning national lung cancer incidence rates were obtained from GLOBOCAN 2012 estimates,22 and data on national prevalence of cigarette smoking was obtained from the World Health Organization Report on the Global Tobacco Epidemic 2013.23Two of the authors (KT and KS) independently extracted study details from the full text articles using a pilot-tested form. When discrepancies arose, a third author (KO) negotiated a consensus. We abstracted data on the number of observed deaths or cases, the number of expected deaths or cases, and the effect measure and the 95% confidence interval (CI) for lung cancer  and bladder cancer (if available). We collected data on bladder cancer to assess heterogeneity between the included studies. For the studies in which CIs were not reported, we calculated them by the exact probabilities of the Poisson distribution using the observed deaths/cases and expected deaths/cases reported in the articles.24 This assessment scale consists of eight items that are categorized into three major components: selection, comparability, and exposure. For cohort studies, a modified version of the NOS19 was used. This modified NOS was developed for assessing the quality of occupational cohort studies and includes five quality components: representativeness of the exposed cohorts, exposure assessment/reporting, comparability of exposed and non-exposed cohort, assessment of outcomes, and adequacy of follow-up.Quality was assessed by two reviewers (KT and KS), and discrepancies were resolved by consulting with a third reviewer (KO). For case-control studies, we used the Newcastle-Ottawa Scale (NOS).25 The presence of between-study heterogeneity was assessed using the I2 test.26 Publication bias was also examined visually using funnel plots and mathematically using the Egger regression asymmetry test.27We performed meta-analysis to obtain the weighted average (pooled) of the SMR and SIR for cohort studies and the OR for case-control studies using the Comprehensive Meta-Analysis version 2.0 software . Effect measures were pooled using random-effects models, which were weighted using the inverse of the variance.In order to explain any heterogeneity seen between studies, we used subgroup analyses. The covariates considered were study outcome , study area (Asia vs Europe vs the United States), reference group , cohort/sample size, type of industry, type of exposure to BZ/BNA, occupational exposure to some carcinogen for lung cancer (no vs yes), year of starting the production/use of BZ/BNA, follow-up duration, SMR/SIR for bladder cancer, national incidence rate for lung cancer, and national prevalence of cigarette smoking. The covariates with continuous variables were divided into two subgroups using the median. High SMR/SIR for bladder cancer was used as the proxy measure for probable higher exposure to BZ/BNA. Based on the information on type of industry, the situation of exposure to BZ/BNA was dichotomized into direct (dye and manufacture) and indirect (leather tanning and rubber).28In our sensitivity analyses, we separately analyzed category outcomes of the assessed study quality variables to determine if there was any relationship between quality and outcome. Additionally, we assessed the influence individual studies had on the pooled estimate by deleting all risk estimates of each study from the meta-analysis and determining the range of pooled estimates. We also conducted a cumulative meta-analysis in publication year order to determine just when the risk estimate became statistically significant and to clarify any variations.29Instead of a dose-response analysis, we did a meta-analysis that combined the results of workers who had the highest occupational BZ/BNA exposure from studies reporting data relevant to level of exposure to BZ/BNA.8–14,30–45 We could identify neither prospective cohort studies nor case-control studies that examined the association between lung cancer and occupational exposure to BZ/BNA. Therefore, all the included studies were retrospective cohort studies. Table Our search identified 460 candidate references Figure . After tn = 9), leather tanning (n = 5), the rubber industry (n = 5), and BZ/BNA manufacturing (n = 4). Regarding smoking, four of the included studies reported information on cigarette smoking, but only one study11 calculated the smoking-adjusted risk for lung cancer. Regarding gender, most of the study subjects were male. Nine studies contained only males, and most cohorts with males and females either excluded females from the analysis or presented risk estimates for males and females combined. Regarding the overall meta-analysis, we used the data for male workers when available. If male data had not been published for the individual studies, the effect estimates for both sexes were used in the overall meta-analysis. For occupational exposure to chemicals other than BZ and BNA  ether,49 which are classified by the IARC as “carcinogenic to humans” (Group 1) based on evidence of increased lung cancer in people.The meta-analysis included studies that covered 1745 cases of lung cancer. Cohort studies had been carried out in Europe, the United States, and Asia from 1976–2015. The industries investigated included dye production  did not use appropriate methods according to the criteria defined in the modified NOS; 4) as for outcome assessments, 19 studies were assessed using formal records ; and 5) regarding follow-up adequacy, over half of the studies (n = 12) had nearly complete follow-up , while 6 studies did not report on loss to follow-up.The results of the study quality assessment are shown in 2 = 64.1%, P < 0.001).A forest plot summarizing the results and weights applied to each study is shown in Figure 2) was the same for both subgroups (I2 = 65.4%). Cohort size, type of industry, situation of exposure to BZ/BNA, and magnitude of the SMR/SIR for bladder cancer were each statistically significant predictors of the pooled risk estimates for lung cancer. Pooled risk estimates were elevated for dyestuff workers  and workers at BZ/BNA manufacturing plants , but not for leather tanning workers  or workers in the rubber industry . Pooled risk estimates were increased for small cohorts  compared with large cohorts ; workers with direct exposure to BZ/BNA  compared with workers with intermittent contact with BZ/BNA ; and cohorts that reported an SMR/SIR for bladder cancer ≥4.7  compared with other cohorts . Pooled risk estimates were similar in workers with occupational exposure to some carcinogens for lung cancer  and workers without occupational exposure to carcinogens for lung cancer . No difference in risk was found between cohorts that started the production/use of BZ/BNA before 1945 and cohorts that initiated it after 1945, as well as between studies with high national incidence rates for lung cancer and studies with low rates. In addition, no difference in risk was shown between cohorts with high prevalence of male smoking and cohorts with low prevalence of male smoking. In subgroup analyses, all but four subgroups  showed significant heterogeneity, with I2 > 50%.There was enough heterogeneity evidence that it was decided to investigate possible explanatory factors. Table Through subgroup analyses, we tried to determine if study quality influenced outcome. The bottom of Table Sensitivity analysis by omitting each study in turn had no effect on the meta-analysis results and showed robust results see . The pooP-value = 0.48).Visual examination of the funnel plot see  used to The results of the cumulative meta-analysis are depicted in Figure 2 = 67.2%, P = 0.009).We identified 6 studies that evaluated the risk estimates of exposure levels among BZ/BNA-exposed workers see . The pooWe conducted a systematic review and meta-analysis to investigate the association between occupational BZ/BNA exposure and risk of lung cancer. A total of 23 retrospective cohort studies of 1745 cases were included in the analysis. We found a significantly increased risk of lung cancer among workers exposed to BZ/BNA occupationally, with an overall pooled risk estimate of 1.28 . Results from cumulative meta-analysis and sensitivity analysis and an absence of publication bias suggest that the results are reliable and robust.29,50 For example, in a meta-analysis of occupational exposure to asbestos and ovarian cancer, the SMR for lung cancer was employed.29 Additionally, workers with direct exposure to BZ/BNA  are likely to have been exposed to higher concentrations of BZ/BNA than workers exposed to BZ/BNA indirectly . Hence, these results provide more evidence that the positive findings in this meta-analysis are due to BZ/BNA exposure. Nevertheless, in addition to higher BZ/BNA exposure, reporting a high risk (risk estimate ≥4.7) of bladder cancer may indicate greater confounding by smoking. Therefore, the possibility of a confounding effect from smoking in the studies identifying increased risks of bladder cancer should be considered.A major finding here is that the pooled risk estimate for lung cancer was elevated both in those studies in which BZ/BNA-associated bladder cancer relative risks were also elevated  and in those studies in which workers had many opportunities for direct exposure to BZ/BNA . The results from this analysis of highly exposed workers also indicate a stronger effect than among all workers combined . In this meta-analysis, no studies reported exposure as work-place measurements, which are the most accurate and unbiased occupational exposure estimates. However, since BZ/BNA exposures are generally apt to be higher in studies finding elevated bladder cancer rates, the presence of a positive bladder cancer finding may be a legitimate substitute for high BZ/BNA exposure. In meta-analysis of cancer risk in occupational settings, a similar approach has been adopted for the purpose of identifying a highly exposed subgroup.51 While some research has found variances in the metabolic patterns of monoarylamines, such as BNA, and diarylamines, such as BZ,52 the main chemical reactions needed in BZ and BNA metabolism are the same; initial activation occurs in the liver, and further metabolism that creates more reactive compounds occurs in the bladder.51,53 This is why bladder cancer is the main effect associated with BZ/BNA exposure. The initial activation of aromatic amines via N-oxidation is mediated primarily by cytochrome P4501A2 (CYP1A2).53 A prior study did not detect CYP1A2 in most human extrahepatic tissue.54 Further metabolism of aromatic amines involves O-acetylation, which is regulated by the enzymes N-acetyltransferases 1 (NAT1).55 NAT1 are detected not only in the human urinary bladder56 but also in the human peripheral lung.57 Although aromatic amine N-oxidation does not take place in the lung, these compounds could undergo N-oxidation in the liver due to CYP1A2, to later circulate to the lung, where they could be O-acetylated by NAT1 found in lung tissue.58 This metabolic activation pathway is supported by prior studies for 4-aminobiphenyl (ABP), which is a prototypical aromatic amine; ABP DNA adducts were detected in human lung tissues obtained by surgery or autopsy,59 as well as in human urinary bladder tissue biopsy samples and exfoliated urothelial cells.60,61 With regard to experimental animals, several studies indicate that exposure to BZ/BNA can increase the incidence of lung cancer; indeed, oral administration of BZ produced a high incidence of lung tumors in mice,62 intraperitoneal injection of BNA increased the incidence of lung adenomas in mice,63 and gavage administration of BNA caused lung tumor multiplicity in mice.64 The capacity of the human lung to metabolically activate BZ and BNA,57 the presence of DNA adducts of an aromatic amine in human lung tissues,59 and the experimental evidence of lung cancer for BZ and BNA in mice62–64 provide biological plausibility for the findings of this meta-analysis.It is not known just how BZ and BNA cause lung cancer, but there is data for a number of possible mechanisms. As with many other aromatic amines, BZ and BNA must be metabolized into reactive electrophiles that react with DNA so they can become carcinogenic.65 Although four studies obtained information on cigarette smoking, two of them calculated smoking-adjusted risk only for bladder cancer,36,38 and one of them had incomplete tobacco smoking data (lacking in about one third of cohort members).10 Only one study11 presented an effect estimate for lung cancer risk from a multivariate model; the relative risk for lung cancer incidence after adjustment for age at first exposure, smoking status, and other occupational co-exposure was 3.02  for long duration of employment compared with the short duration group. Lack of adjustment for smoking might lead to overestimation of the association between lung cancer and exposure to BZ/BNA. Therefore, further studies with sufficient data on smoking are required for a better understanding of the risk for lung cancer in relation to occupational exposure to BZ/BNA.A major concern in interpreting our analysis was its inability to account for smoking, which is an important non-occupational risk factor for lung cancer.66 In observational epidemiology, study designs, populations, exposure assessing and outcome methods, and statistical analyses are rarely, if ever, the same.50 Therefore, meta-analysis of observational studies has the potential to introduce bias and confounding.15 Although we conducted subgroup analysis to explore sources of heterogeneity, study characteristics and study quality had little influence on our heterogeneity. A prior review based on 60 meta-analyses in occupational epidemiology pointed out that the most obvious finding was the various exposure definitions these studies employed and the lack of exposure data, which make comparisons onerous.67 The studies included in this meta-analysis also used a variety of BZ/BNA exposure surrogates, such as occupational groups, years of employment, and duration of employment. The differences in definitions of BZ/BNA exposure across studies might explain some of the heterogeneity between studies.A further limitation of our findings is the issue of heterogeneity between studies. Statistically significant heterogeneity was noted not only in the meta-analysis of all studies combined but also in subgroup analysis and sensitivity analysis. Heterogeneity is due to methodological diversity between the studies and shows up in the observed effects, being even far more different from each other than one would expect randomly.We are aware that occupational cohorts included in this study not only had simple exposures to BZ/BNA but had exposures to other chemicals as well. Although we conducted subgroup analyses and confirmed no difference in the lung cancer risk between workers with and without occupational exposure to known lung carcinogens, concomitant occupational exposures to other chemicals was not taken into consideration. Therefore, the results of this meta-analysis might be biased due to exposure to other chemicals.68 However, a main drawback of case-control studies is that exposure information is obtained by self-reporting after disease has occurred.68 Differential misclassification of exposure due to recall bias is likely, tending toward overestimation of the association between exposure and outcome. Therefore, including no case-control studies might prevent overestimation of the risk for lung cancer among workers exposed to BZ/BNA.Even though we systematically searched for all published cohort and case-control studies indexed in MEDLINE, EMBASE, and CINAHL, none of the identified studies were case-control studies. In general, case-control studies are superior to retrospective cohort studies regarding adequate control of confounding factors, including smoking.18; 2) the total number of lung cancer deaths or incident cases was 1745 from all the cohort studies, providing summarized epidemiological evidence with adequate statistical power to examine the association between BZ/BNA exposure and lung cancer; and 3) our meta-analysis, restricted to workers with probable higher exposure, is reasonably consistent with an underlying dose-response effect.Despite these limitations, this study has the following strengths: 1) this is the first meta-analyses of lung cancer risk in workers exposed to BZ/BNA conducted according to the method of prospective meta-analysis, which means a meta-analysis of studies that were identified, evaluated, and determined to be eligible for the meta-analysis before the results of any of those studies became knownIn conclusion, this systematic review and meta-analysis suggests that occupational BZ/BNA exposure is associated with an increased risk of lung cancer. One important finding of this study is that pooled risk estimates were raised in the subgroup of studies in which workers were probably exposed to a high level of BZ/BNA. Although confounding by smoking cannot be completely ruled out, and the heterogeneity among studies requires cautious interpretation, our findings suggest that a lung cancer diagnosis following occupational BZ/BNA exposure may need to be considered a potential occupational disease.eTable 1. eTable 2. eTable 3. eTable 4. eTable 5. eTable 6. eFigure 1. eFigure 2. Abstract in Japanese.