ADIPOQ, triggering this up-to-date meta-analysis.Inconsistencies have existed in research findings on the association between cardiovascular disease (CVD) and single nucleotide polymorphisms (SNPs) of We searched for relevant studies in PubMed, EMBASE, Cochrane Library, CNKI, CBM, VIP, and WanFang databases up to 1st July 2017. We included 19,106 cases and 31,629 controls from 65 published articles in this meta-analysis. STATA 12.0 software was used for all statistical analyses.Our results showed that rs266729 polymorphism was associated with the increased risk of CVD in dominant model or in heterozygote model; rs2241766 polymorphism was associated with the increased risk of CVD in the genetic models . In subgroup analysis, significant associations were found in different subgroups with the three SNPs. Meta-regression and subgroup analysis showed that heterogeneity might be explained by other confounding factors. Sensitivity analysis revealed that the results of our meta-analysis were stable and robust. In addition, the results of trial sequential analysis showed that evidences of our results are sufficient to reach concrete conclusions.In conclusion, our meta-analysis found significant increased CVD risk is associated with rs266729 and rs2241766, but not associated with rs1501299.The online version of this article (10.1186/s12944-018-0767-8) contains supplementary material, which is available to authorized users. Cardiovascular disease (CVD) is the primary cause of death worldwide, leading to 32% of all deaths worldwide in 2013 . IdentifAdiponectin is involved in CVD: low levels of adiponectin (hypoadipoectinemia) positively correlate with the risk of CVD, and higher levels of adiponectin protect against this disease –11. AdipADIPOQ which is located in chromosome 3q27 [ADIPOQ [ADIPOQ have been found to be associated with CVD [ADIPOQ  and CVD/subclinical CVD [ADIPOQ  were most widely studied. However, findings from previous studies on the three SNPs in relation to CVD risk are inconsistent and inconclusive.Adiponectin is encoded by ome 3q27 , and adi [ADIPOQ . SNPs inwith CVD , 24, diawith CVD , 26, strwith CVD , myocardwith CVD , 30, canwith CVD , 32, kidwith CVD , 34, andwith CVD , 36. Preical CVD . The thrADIPOQ, Du et al. [For rs266729  in u et al.  and Zhanu et al.  found thu et al.  revealedu et al. , Cheong u et al. , and Chiu et al.  and Jungu et al.  identifiu et al. , Oliveiru et al. , and Mofu et al.  found thu et al.  revealedu et al.  and Esteu et al.  revealedu et al. , howeveru et al.  found thADIPOQ and CVD were significant but weak [ADIPOQ and susceptibility to CVD [ADIPOQ and CVD in order to get a more precise and reliable assessment of the association.Meta-analysis performed by Zhang et al. in 2012 revealed that associations between the SNPs  in y to CVD , 38, 45.ADIPOQ polymorphisms and CVD risk up to July 1st, 2017. The literature search was done without any language or population restrictions imposed. During the literature search, we used various combinations of keywords, such as ‘coronary heart disease (CHD)’ or ‘coronary artery disease’ or ‘cardiovascular disease’ or ‘ischemic heart disease’ or ‘angina’ or ‘myocardial infarction (MI)’ or ‘stroke’ or ‘atherosclerosis’ or ‘arteriosclerosis’ or ‘coronary stenosis’ combined with ‘ADIPOQ’ or ‘APM1’ or ‘ACDC’ or ‘adiponectin gene’ and ‘polymorphisms’ or ‘variants’ or ‘variations’. Joseph Sam Kanu and Shuang Qiu independently performed the literature search for potential articles included in this meta-analysis. All articles retrieved were first organized in reference manager software (Endnote 6).We performed an extensive literature search in PubMed, EMBASE, Cochrane Library, CNKI, CBM, VIP, and WanFang databases for published articles on the association between ADIPOQ SNP to be assessed; 2) the study included original data (independence among studies); 3) evaluation of the ADIPOQ polymorphisms  and CVD risk; 4) the language of the study was English or Chinese; and 5) observed genotype frequencies in controls must be consistent with Hardy–Weinberg equilibrium (HWE). We excluded a study based on: 1) the study contained overlapping data; 2) the study with missing information (particularly genotype distributions), after corresponding author, who was contacted by us with email, failed to provide the required information; and 3) genome scans investigating linkages with no detailed genotype distributions between cases and controls. Where there was a disagreement on the selection of a study, the issue was resolved by discussion or consensus with the third investigator (Ri Li). For articles with missing data, we emailed the corresponding authors for the required data.A study included in this meta-analysis was based on the following criteria: 1) the study has sufficient data to allow association between CVD risk and We used the NATURE-published guidelines proposed by the NCI-NHGRI Working Group on Replication in Association Studies for assessing the quality of each study included in this meta-analysis . These gJoseph Sam Kanu and Shuang Qiu extracted data from each study independently. We summarized the information extracted from each article in Table P < 0.05 was considered as a significant deviation from HWE. The strength of association between the three ADIPOQ polymorphisms and CVD susceptibility was assessed using odds ratios (OR) and 95% confidence intervals (95% CI). The associations were measured based on five different genetic models: allelic model , dominant model , recessive model , heterozygote model , and homozygote model . Heterogeneity were evaluated by the Chi-square test based Q-statistic, and quantified by I2-statistic [P > 0.10, I2 ≤ 50%), data were pooled by fixed-effect model (Mantel and Haenszel methods); otherwise, the heterogeneity was evaluated by random-effect model (DerSimonian and Laird methods). Meta-regression analysis was performed to detect main sources of heterogeneity. In addition, subgroup analyses were stratified by population , genotyping method , sample size (< 1000 and ≥ 1000), and quality score (< 10 and ≥ 10). Sensitivity analysis was performed to examine stability of our results by omitting each study in each turn. Publication bias was measured by funnel plots [P < 0.05 considered statistically significant publication bias). STATA 12.0 software  was used for all statistical analyses. P-value < 0.05 was considered statistically significant, except where other-wise specified. A separate analysis was performed for each SNPs included in the meta-analysis.HWE was evaluated for each study using Goodness of fit Chi-square test in control groups, and tatistic . If therel plots , and quael plots  . Thus, we selected random-effect model, and found that rs266729 polymorphism was associated with the increased risk of CVD in dominant model  and in heterozygote model   polymorphism and CVD included 29 studies with 29,021 subjects . Significant heterogeneity among studies was observed  and under heterozygote model  in East Asian. On the basis of genotyping methods, a significant risk association between rs266729 polymorphism and CVD was found when genotyping was performed using PCR-RFLP method under dominant model  and under heterozygote model . On the basis of sample size or quality score, we found that rs266729 polymorphism was associated with the increased risk of CVD under allelic, dominant, and heterozygote models , after pooled the ORs by the subgroups of sample size ≥ 1000 or quality score ≤ 10 , we found that rs2241766 polymorphism was associated with the increased risk of CVD in the five genetic models    polymorphism and CVD included 40 studies with 25,548 subjects . Using inverse-variance weighted random effect model (5) Table , Fig. 3.OR > 1 and P < 0.05). Secondly, on the basis of genotyping method, the results that genotyping was done by PCR-RFLP or other methods showed that rs2241766 polymorphism was associated with the increased risk of CVD under five genetic models . Thirdly, on the basis of sample size, rs2241766 polymorphism was associated with the increased risk of CVD under the five genetic models in the subgroup of sample size ≤1000 , but was associated with the decreased risk of CVD in the subgroup of sample size ≥1000 under recessive model  and under homozygote model . Finally, on the basis of quality score, when we pooled the ORs by the subgroups of quality score ≤ 10, we found that rs2241766 polymorphism was associated with the increased risk of CVD under the five genetic models  , we found that there was no association between rs1501299 polymorphism and CVD in the five genetic models   polymorphism and CVD included 44 studies with 37,371 subjects . Using the inverse-variance weighted random effect model (P > 0.05)  Table .P = 0.019; dominant model: P = 0.032; recessive model: P < 0.001; and homozygote model: P < 0.001) and quality score  contributed to the observed heterogeneity across all the studies of the association between rs2241766 polymorphisms and CVD risk. However, in the meta-analysis of the associations between rs266729/rs1501299 polymorphisms and CVD risk, we did not identify the source of heterogeneity  . Publication biases were found in analyses of the associations between rs2241766 polymorphisms and CVD risk , and between rs1501299 polymorphisms and CVD risk   to investigate the association between In view of the association between rs266729 and CVD risk, Yang et al. (2012) , Zhou etThe association between rs2241766 and CVD risk also has been the subject of meta-analysis –64. ThesZhou et al. revealed no significant association between rs1501299 polymorphism with CAD susceptibility [With regard to the association between 1,501,299 and CVD, the results are also conflicting , 61. Zhotibility . Zhang etibility . In compADIPOQ polymorphisms on adiponectin levels [ADIPOQ polymorphism and CVD risk may be due to differences in publication bias, sample size, or insufficient statistical power. In addition, evidences have showed that studies which deviate from HWE in controls may reflect the presence of genotyping errors, population stratification, and selection bias in the controls (or without representation of studied sample). Thus, including those studies may decrease the quality of a meta-analysis or generate inconsistent results [Different genetic admixture and environmental factors among human populations, which tend to explain ethnic background, strongly modulate the effects of n levels , 66. Stun levels . These c results .Heterogeneity across all the studies of the associations should be noted because it may potentially affect the strengths of the present meta-analysis. We, thus, used random effect model. Our results showed that sample size and quality score are the factors of heterogeneity across all studies of association between rs2241766 polymorphisms and CVD, but no factors contribute the heterogeneity across all studies of association between rs266729/rs1501299 polymorphisms and CVD. However, heterogeneity was still high in the subgroup analysis of the two factors. For these reasons, heterogeneity might be explained by other confounding factors, such as gene-gene interaction and gene-environment interaction.Our meta-analysis has some limitations. Firstly, significant publication bias was found in the analysis of rs2241766  and rs1501299 (under recessive model). Secondly, our meta-analysis mainly included Europeans and Asians with only few other races, thus limiting our power to generalize our findings in other races. Finally, our results might be affected by the potential weaknesses of genetic association studies, such as phenotype misclassifications, genotyping error, population stratification, gene-environment or gene-gene interactions, and selective reporting biases , 69.ADIPOQ variant and CVD, which give our study more statistical power to draw valid conclusion on this issue. Secondly, sensitivity analysis showed that the results of our meta-analysis are stable and robust. Thirdly, the evidence of our results are sufficient to reach concrete conclusions, which were proved by TSA for the first time. We strongly believe our findings will help settle some of the controversies surrounding the ADIPOQ-CVD association research.Despite the limitations highlighted above, our meta-analysis also had some strength. Firstly, we searched extensively and investigated more studies and more participants than any other meta-analyses performed on the association between ADIPOQ variants and CVD.Our meta-analysis found significant increased CVD risk is associated with rs266729 and rs2241766, but not associated with rs1501299. Investigating gene–gene and gene–environment interactions is needed to give more insight into the genetic association between Additional file 1:Summary of three SNPs characteristics. (DOCX 48 kb)Additional file 2:Meta-regression results of the association between the SNPs and CVD risk. (DOCX 59 kb)Additional file 3:Funnel plots of three SNPs for publication bias. (DOCX 1250 kb)Additional file 4:Additional references. (DOCX 22 kb)


LIM domain only 1 (LMO1) gene polymorphisms affect neuroblastoma susceptibility. In this work, we used Taqman methodology to genotype four LMO1 polymorphisms  in 118 neuroblastoma cases and 281 controls from Northern China. Odds ratios (ORs) and 95% confidence intervals (CIs) were used to evaluate the association. We found that rs4758051 G > A was associated with a decreased neuroblastoma risk . Likewise, carrying the rs10840002 G allele was also associated with a decreased neuroblastoma risk in this Northern Chinese population. In a combination analysis using Southern and Northern Chinese populations, we found that those carrying the rs110419 G, rs4758051 A or rs10840002 G allele were at decreased neuroblastoma risk, and this finding was supported by a false-positive report probability analysis. These results further verify that LMO1 polymorphisms are protective against neuroblastoma. Case-control studies with larger samples and using other ethnicities are still needed to confirm our conclusion.Previous genome-wide association and validation studies suggest that  Neuroblastoma is a common solid tumor derived from primordial sympathetic neural precursors and has complicated clinical manifestations . Around The etiology of neuroblastoma is not yet fully understood, and the risk factors affecting the susceptibility to neuroblastoma have not been well documented . EpidemiLIM domain only 1 (LMO1) gene is located on chromosome 11p15 and encodes an intertwining LIM-only transcriptional regulator [LMO1 is a member of the LMO gene family [LMO1 gene single nucleotide polymorphisms (SNPs) reportedly affect susceptibility to acute lymphoblastic leukemia [The egulator . LMO1 ise family , 18 and e family , though e family . LMO1 prleukemia  and neurleukemia .LMO1 polymorphisms and neuroblastoma risk in a Southern Chinese population [LMO1 polymorphisms and neuroblastoma risk, we performed the current hospital-based case-control study using subjects from Northern China.We previously investigated the association between pulation . Given tP = 0.189) or gender (P = 0.196) distribution between cases and controls. According to the INSS criteria [The demographic characteristics of the included 118 cases and 281 controls are summarized in criteria , 15 (12.LMO1 polymorphisms and their associations with neuroblastoma risk are listed in Table PHWE > 0.05). Moreover, the analyses indicated that carrying the rs4758051 A allele had a protective effect against developing neuroblastoma (AA vs. GG: adjusted odds ratio (OR) = 0.28, 95% confidence interval (CI) = 0.13–0.62, P = 0.002; AG/AA vs. GG: adjusted OR = 0.62, 95% CI = 0.40–0.97, P = 0.035; AA vs. GG/AG: adjusted OR = 0.33, 95% CI = 0.15–0.69, P = 0.003). Similarly, we found that carrying the rs10840002 G allele was associated with a decreased risk of neuroblastoma . However, we failed to detect an association between the rs110419 A > G or rs204938 A > G polymorphism and neuroblastoma risk, whether or not adjusted for age and sex.The genotype frequencies of the four LMO1 polymorphisms and neuroblastoma risk, we combined our present results the data from our earlier study [P = 0.004; GG vs. AA: adjusted OR = 0.58, 95% CI = 0.40–0.84, P = 0.004; AG/GG vs. AA: adjusted OR = 0.65, 95% CI = 0.50–0.83, P = 0.001), as were those carrying the rs4758051 A allele  or the rs10840002 G allele . No significant association was observed between rs204938 A > G and neuroblastoma risk.To further elucidate the association between er study . In the False-positive report probability (FPRP) analysis showed that all of the statistically significant associations remained noteworthy, when a prior probability of association of 0.25 was considered. At a prior probability level of 0.1, all except one association (rs10840002 A > G) remained noteworthy. At a prior probability level of 0.01, only the association between rs110419 A > G and neuroblastoma risk remained noteworthy (FPRP = 0.168). Detailed information from the FPRP analysis is listed in Table LMO1 polymorphisms on neuroblastoma risk in a Northern Chinese population. Consistent with our earlier findings [LMO1 polymorphisms were associated with a decreased risk of neuroblastoma. To our knowledge, this is the first study to validate the association between LMO1 polymorphisms and neuroblastoma risk using two resident groups in China. This combined analysis improves the statistical power for assessing the impact of LMO1 polymorphisms on neuroblastoma risk.In the present case-control study, we further verified the effect of findings , we obseLMO1 polymorphisms were associated with predisposition to neuroblastoma [LMO1 gene  that were associated with neuroblastoma risk in subjects of European ancestry. Thereafter, case-control studies conducted with Italian [LMO1 polymorphisms to be factors affecting neuroblastoma risk. In 2016, we conducted the first epidemiological study on the effect of LMO1 polymorphisms on neuroblastoma susceptibility in a Southern Chinese population [LMO1 SNPs in 256 cases and 531 controls. We only found that the rs110419 A > G polymorphism was associated with a significantly lower neuroblastoma risk. However the strength of that conclusion was limited by the small sample. We have therefore now expanded the size of our sample.A previous genome-wide association study revealed that blastoma . In that Italian , African Italian  and Nort Italian  populatipulation . We genoLMO1 polymorphisms and neuroblastoma risk in an additional 118 cases and 281 controls. Our analysis indicates that carrying the rs4758051 A or rs10840002 G allele is associated with decreased risk of neuroblastoma in a Northern Chinese population. In our earlier study of a Southern Chinese population, we detected a significant protective association between only the rs110419 A > G polymorphism and neuroblastoma risk [Here, we assessed the relationship between oma risk . Two posLMO1 polymorphisms on neuroblastoma risk. In addition, the FPRP analysis also enhances the robustness of our findings.To increase the representation for our conclusions, we combined the results from our present study with those from our earlier one. The combined analysis indicated that carrying the rs110419 G, rs4758051 A or rs10840002 G allele was associated with a decreased risk for neuroblastoma. The larger sample in the combined analysis highlights the important protective effect of LMO1 polymorphisms and neuroblastoma risk in the Chinese, several limitations still exist. First, we only genotyped four SNPs in this study, other potentially functional polymorphisms not discovered in genome-wide association studies were omitted. These include the rs2168101 G > T polymorphism, which was recently found to be associated with neuroblastoma [LMO1 polymorphisms is needed.Although this is a relatively large sample for investigating the correlation between blastoma . Second,LMO1 polymorphisms on neuroblastoma susceptibility in a Chinese population. More case-control studies based on other ethnicities and multicenter investigations are encouraged to support these observations.In summary, we have further confirmed the protective effect of The recruitment criteria for neuroblastoma patients and controls were described previously , 27, 28.LMO1 SNPs  identified as being associated with neuroblastoma in an earlier genome-wide association study were selected  was used to perform all statistical analyses.The goodness-of-fit χanalysis , 34. We


PRSS1 p.R122H mutation was the first discovered to affect hereditary CP, with 80% penetrance. We performed here a systematic review and meta-analysis to evaluate the associations of PRSS1 p.R122H mutation with CP of diverse etiology.  Environmental factors and genetic mutations have been increasingly recognized as risk factors for chronic pancreatitis (CP). The  The PubMed, EMBASE, and MEDLINE database were reviewed. The pooled odds ratio (OR) with 95% confidence intervals was used to evaluate the association of p.R122H mutation with CP. Initial analysis was conducted with all etiologies of CP, followed by a subgroup analysis for hereditary and nonhereditary CP, including alcoholic or idiopathic CP. PRSS1 p.R122H mutation was significantly associated with an increased risk of CP (OR = 4.78[1.13–20.20]). Further analysis showed p.R122H mutation strongly associated with the increased risk of hereditary CP (OR = 65.52[9.09–472.48]) but not with nonhereditary CP, both alcoholic and idiopathic CP.  A total of eight case-control studies (1733 cases and 2415 controls) were identified and included. Overall,  Our study showing the differential role of p.R122H mutation in various etiologies of CP indicates that this complex disorder is likely influenced by multiple genetic factors as well as environmental factors. The  factors . The rolPRSS1 mutation was first discovered to be associated with the phenotype of hereditary pancreatitis (HP) twenty years ago [PRSS1 gene, detected in more than 50% of the patients with HP, was the most common mutation in patients with HP [PRSS1 mutant p.R122H transgene exhibited early onset of acinar cell injury and inflammation that progressed with age to a chronic stage and associated with fibrosis and acinar cell differentiation [PRSS1 mutants (p.R122H or p.N29I) in murine pancreatic acinar cells also promoted pancreatic injury [ears ago . Carrierears ago , 9. R122 with HP . A singl with HP , 11. Micc injury . Geneticc injury , but exhc injury .PRSS1 gene and the risk of CP. We used PRSS1 p.R122H as an example of genetic risk factor in order to understand more about the association of single genetic factor in CP patients. We found that PRSS1 p.R122H mutation was significantly associated with an increased risk of hereditary CP. Further analysis suggested p.R122H mutation was weakly but still significantly associated with an increased risk of nonhereditary CP, including alcoholic or idiopathic CP with a slightly greater risk in idiopathic CP.In this study, we performed a systematic review and meta-analysis to investigate the association between p.R122H mutation in PRSS1 or p.R122H were combined with pancreatitis. The filter “human” was applied. The search was conducted using database entries from January 1996 to July 2016. The language was limited to English. The duplicates were removed manually. We also reviewed reference list of citations in the identified publications for additional citations.A search of the literature was conducted by two authors (C.H and L.W) independently using PubMed, EMBASE, and MEDLINE electronic databases. The keywords n = 36). All basic and animal experimental studies were also excluded (n = 26). Review articles (n = 67) and case reports (n = 39) were excluded. Studies that investigated autoimmune pancreatitis (n = 1) or recurrent acute pancreatitis (n = 3) were also excluded.The published observational studies found by our search that reported the association of chronic pancreatitis with p.R122H mutation were included in this review. Studies that enrolled pediatric patients were excluded (Data were extracted by two authors (C. H and L. W) independently for each individual study. The following data were extracted: year, study design, racial background, genotyping method, criteria used for diagnosis of CP, number of patients diagnosed as CP, total number of patients, etiology of CP , history of alcohol consumption, and history of smoking if applicable.The quality of included case-control studies was systematically assessed using the Newcastle Ottawa Scale (NOS). A maximum of 9 points were given to each study within the categories of the selection, comparability, and exposure, as scored by two independent observers. Studies were considered to be “good” quality if they scored >5 points, whereas studies were considered to be “poor” quality if they received ≤5 points.The American Pancreatic Association (APA) guidelines  were useAlcohol was considered the cause of CP in those who consumed more than 60 or 80 g per day for at least 2 years. Smokers were defined as those who smoked 10 or more cigarettes per day for at least 2 years.PRSS1 p.R122H mutation and the risk of CP was initially evaluated. A preliminary meta-analysis combining all studies included regardless of etiology was conducted. Further stratified analysis was used to assess the risk of hereditary and nonhereditary CP with PRSS1 p.R122H mutation, including subanalysis of alcoholic and idiopathic CP. Forest plots were created showing the pooled odds ratio (OR) with the corresponding 95% confidence intervals (CIs). Data were pooled using the random effects model to provide a more conservative estimate. Statistical heterogeneity between studies was determined with Cochran Q test and the I2 value [p value of ≤0.05 was considered as statistical significance. Sensitivity analysis was used to test the robustness of associations by omitting each individual study in turn from all available studies when appropriate. Publication bias was assessed by visual inspection of funnel plot [For each study, the association between I2 value . High stI2 value . A p valnel plot . The anan = 36), animal and experimental studies (n = 28), case reports (n = 39), cross-sectional studies (n = 1), review (n = 67), editorial (n = 2), and others (n = 12). After removing, seventy-three articles potentially met the inclusion criteria for eligibility, and 63 were excluded for the reasons as follows: not relevant (n = 46), no genotyping for PRSS1 (n = 7), no genotyping for p.R122H (n = 6), recurrent AP (n = 3), and autoimmune pancreatitis (n = 1). Overall, ten studies [https://doi.org/10.1155/2017/9505460.A total of 567 studies were retrieved from PubMed, EMBASE, and MEDLINE. After removal of duplicates, 258 studies were screened by the title and abstracts. In the preliminary screening, we excluded 185 studies because of the following reasons: pediatric patients .Initial meta-analysis of individual studies and the combined results of the eight studies with all etiologies of CP are shown in p = 0.235 and I2 = 29.5%). The pooled OR revealed an association between p.R122H mutation and the risk of hereditary CP of 65.52 (9.09–472.48). Eight studies evaluated patients with nonhereditary CP . The pooled OR revealed an association between p.R122H mutation, and an increased risk of nonhereditary CP is 2.79 . These data indicate that p.R122H mutation is strongly associated with hereditary CP and has a weak association with nonhereditary CP.Four of the studies assessed patients with hereditary CP . The pooled OR of 3.39 (0.79–14.54) revealed a trend that p.R122H mutation may be associated with an increased risk of alcoholic CP. Since there is only one study with a small size suggesting an association between p.R122H mutation and alcoholic CP and 95%CI crossed 1, we think the p.R122H mutation may not be associated with alcoholic CP, although further study is required to confirm these findings.Six studies assessed patients with alcoholic CP . The pooled OR is 4.43 (1.03–19.05), which revealed a significant weak association between p.R122H mutation and an increased risk of idiopathic CP. Consistent with the analysis in nonhereditary CP, there was a weak but still significant association between p.R122H mutation and an increased risk of CP.Seven studies assessed patients with idiopathic CP , indicating that our results are relatively stable and credible.PRSS1 p.R122H mutation is significantly associated with an increased risk of CP. Our finding is consistent with the meta-analysis conducted by Liu and Zhang [PRSS1 gene. But this study neither specified the sites of the gene mutations nor compared the different pancreatitis etiologies. In our current study, we specifically asked whether there is an association between PRSS1 p. R122H and chronic pancreatitis with subgroup analysis applied to the known etiologies. Further analysis showed that p.R122H mutation is strongly associated with hereditary CP. Since the first report in 1996 indicating that PRSS1 p.R122H mutation was associated with the phenotype of HP in a North American family [PRSS1 gene and HP have been reported in a number of studies from many parts of the world including Europe and Asia [Overall, we found that the nd Zhang  which sun family , strong and Asia –31. Subgand Asia , 33. We PRSS1 gene mutations in exon 2 and exon 3 have been reported to be pathogenic and associated with the phenotype of CP. p.R122H and p.N29I mutations are the most common mutations [PRSS1-PRSS2 locus account for an increased risk for alcoholic and sporadic CP, but rare PRSS1 gene variants, including three known diseases that were also screened and identified, were not associated with the observed phenotype [Various utations . Recentlhenotype . This fihenotype . Both stPRSS1 p.R122H mutation is significantly associated with an increased risk of CP overall. The differential role of PRSS1 p.R122H mutation was observed in both hereditary and nonhereditary CP. These data suggest that multiple genetic and environmental factors determine the initiation and progression of CP.In summary, our findings indicate that the Supplementary Table 1 Methodologic Quality assessment of included studies according to the Newcastle-Ottawa Scale. Supplementary Figure 1 Sensitivity analysis on the studies.


P < 0.001 for heterogeneity; I2 = 79.6%). There was no significant publication bias by Begg's test (P = 0.348) or Egger's test (P = 0.130). In conclusion, this meta-analysis indicates that high intake of carrot is associated with a low incidence of urothelial cancer. Considering the limited included studies and huge heterogeneity, further large well-designed prospective cohort studies are warranted to confirm the findings from our meta-analysis.Previous studies regarding the relationship between carrot intake and risk of urothelial cancer have reported conflicting results. Hence we performed a meta-analysis of eligible studies to summarize evidence on this association. A comprehensive search up to January 2017 was performed in PubMed, Web of Science, Scopus, EMBASE, Cochrane register, and Chinese National Knowledge Infrastructure (CNKI) databases. The combined odds ratio (OR) with 95% confidence interval (CI) for the highest versus the lowest intake of carrot was calculated. A total of six epidemiological studies consisting of four case-control and two cohort studies were included. Overall analysis indicated a significantly reduced risk of urothelial cancer for high intake of carrot . Obvious significant heterogeneity was observed among included studies ( Urothelial cancer is the second most common cancer of the genitourinary tract . UrothelIn epidemiological studies, such as case-control and prospective cohort studies, the potential relationship between carrot consumption and site-specific cancer incidence has been investigated. A dose-response meta-analysis indicated that carrot intake might be associated with a reduced risk of prostate cancer . FallahzThe aim of this study was to evaluate the potential relationship between carrot intake and urothelial cancer risk by performing a meta-analysis of all eligible case-control and cohort studies. We also carried out stratified meta-analysis based on study design, geographical region, gender, and carrot type.n = 3), Asia (n = 2), and USA (n = 1). All included studies were published between 1979 and 2005, of which two were cohort and four were case-control studies. Information on carrot intake was collected by face-to-face interview or a self-administered questionnaire. Table The detailed steps of literature search were present in Figure  USA n = . All incP < 0.001 for heterogeneity; 2I = 79.6%).Figure We then conducted subgroup analyses by study design, study region, carrot type, and gender Table . In the In the influence analysis, the impact of each study on the combined OR was checked by repeating the meta-analysis after omission of each study in turn. As shown in Figure P = 0.348) or Egger's test (P = 0.130).There was no significant publication bias by Begg's test . Heterogeneity is often a concern in a systematic review and meta-analysis, due to the variation in study design, sources of study population, sample size, definitions of carrot intake, and so on. Therefore, we performed several stratified meta-analysis based on study design, geographical region, gender, and carrot type. However, obvious heterogeneity was still observed in most subgroups, which indicated that the heterogeneity can't be explained by a single factor.In this study, there was statistically significant heterogeneity among included studies , while the others used tertiles or grams/day. Furthermore, the species of carrot also varied among included studies. Finally, residual or unknown confounders cannot be completely excluded as a potential bias. Some other special influential lifestyle might also impact the current analysis.In conclusion, this meta-analysis indicates that high intake of carrot is associated with a low incidence of urothelial cancer. Moderate consumption of carrot has been recommended by many researchers, as carrot intake has been reported to be associated a low incidence of several cancers , 7 and pWe performed a comprehensive search in PubMed, Web of Science, Scopus, EMBASE, Cochrane register, and Chinese National Knowledge Infrastructure (CNKI) databases from their inception to January 2017, using the following search algorithm: (diet or nutrition or vegetable or vegetables or carrot or carrots or carotenoids) and . We assessed potentially relevant studies by screening their titles and abstracts. Full texts for articles matching the eligible criteria were retrieved. We also examined the cited references from retrieved articles and reviews to identify any additional relevant studies. There was no language limitation. This meta-analysis follows the standards of quality for reporting systematic review and meta-analysis .Studies included in this meta-analysis met all the following criteria: (i) evaluated the relationship between carrot intake and urothelial cancer risk, (ii) had a case-control, nested case-control study, or cohort study design, (iii) reported risk estimate and its 95% CI. If multiple studies from the same general population were available, the largest and most detailed study was included in this meta-analysis .Data were extracted independently by two reviewers (XL and HL) with a predefined data collection form. Disagreement was resolved by discussing with a third reviewer (YL). For each study, the following information were collected: first author's surname, publication year, the country in which the study was performed, study design, number of cases, type of carrots, mostly adjusted risk estimates for highest versus lowest level of carrot intake, and matched or adjusted confounders. Considering that urothelial cancer is a rare disease, the relative risk (RR) was assumed approximately the same as OR, and the OR was used as the study outcome. Any adjusted ORs with 95% CIs were extracted directly from the original reports or calculated indirectly with the available data.2I score [P value for heterogeneity was < 0.10 or 2I was > 50%. In this study, the pooled ORs with 95% CIs were estimated with a DerSimonian and Laird random-effects model [P-values.To compute a combined OR with its 95% CI, we extracted the mostly adjusted risk estimates provided in each included study. Homogeneity of ORs across studies was evaluated by Q statistic and the I2 score . The nults model , which tts model  and Eggets model . All of


Aim: The aim of this study was to perform a comprehensive meta-analysis of the association between coffee consumption and risk of endometrial cancer. Methods: Eligible studies were identified by searching the PubMed and EMBASE databases. The dose–response relationship as well as the risk of endometrial cancer for the highest versus the lowest categories of coffee consumption were assessed. Subgroup analyses considering the menopausal and receptor statuses, the smoking status, and the BMI (Body Mass Index) were performed in order to identify potential confounders. Results: We identified a total of 12 studies eligible for meta-analysis. A dose–response meta-analysis showed a decreased risk of endometrial cancer. Moreover, a subgroup analysis indicated that coffee consumption is significantly associated with a decreased risk of postmenopausal cancer. Increasing coffee consumption by four cups per day was associated with a 20% reduction in endometrial cancer risk (relative risk (RR) 0.80; 95% confidence interval (CI) 0.72 to 0.89) and with a 24% reduction in postmenopausal cancer risk . Conclusions: Our findings suggest that increased coffee consumption is associated with decreased risk of endometrial cancer, and this association is observed also for postmenopausal cancer. Endometrial cancer is the third most common female cancer, after breast cancer and cervical cancer: in 2012, worldwide, there were over 1.2 million women with a diagnosis of endometrial cancer made in the previous 5 years. Similarly, in Europe there were 370,000 women with the same diagnosis . Among oA decreased risk of endometrial cancer in women who regularly drink coffee has been documented in retrospective and prospective studies: Je and Giovannucci performed a meta-analysis including 10 case-control and 6 cohort studies and found a significant inverse association of endometrial cancer with coffee consumption in the highest versus the lowest category of coffee intake . The dosWe followed the Meta-Analysis of Observational Studies in Epidemiology (MOOSE) protocols throughout the design, execution, analysis, and reporting of the current meta-analysis  20]..20].http://www.ncbi.nlm.nih.gov/pubmed/) and EMBASE (http://www.embase.com/); the databases were screened from the earliest available online indexing year up to March 2017, with English-language restriction. We included the following search terms: (coffee OR caffeine OR beverages OR diet Or dietary) AND  AND (cancer OR carcinoma OR neoplasm)  . Two autA standardized extraction form was used to extract data from each study included in the meta-analysis. The following information was collected: (1) first author name; (2) year of publication; (3) study cohort name; (4) country; (5) number of participants; (6) sex of participants; (7) age range of the study population at baseline; (8) categories of consumption; (9) follow-up period; (10) endpoints and cases; (11) distributions of cases and person-years, RRs, HRs, and 95% CIs  for all categories of exposure; (12) covariates used in adjustments. Two authors independently performed such process, and discrepancies were discussed and resolved by consensus. Each study was critically appraised through the use of a quality scale, namely, the Newcastle-Ottawa Quality Assessment Scale , which cI2 statistic. The level of significance was set equal to 0.10 for the Q test. The I2 statistic explains the amount of total variation that could be attributed to heterogeneity. I2 values ≤25%, 25–50%, 50–75%, and >75% indicated no, small, moderate, and significant heterogeneity, respectively. To evaluate the stability of the results and potential sources of heterogeneity, a sensitivity analysis by exclusion of one study at a time was performed. Moreover, a subgroup analysis was performed in order to check for potential sources of heterogeneity according to the geographical area. To test for potential confounders and effect modifiers, other subgroup analyses were performed . The publication bias was evaluated by a visual investigation of the funnel plots for potential asymmetry.In this meta-analysis, HRs were deemed equivalent to relative risks (RRs) . For eacp-value for non-linearity by testing the coefficient of the second spline as equal to zero. We performed all analyses with R software version 3.0.3 .To better explore the relationship between exposure and outcome, a dose–response meta-analysis was performed . Extracted data were stratified by the level of coffee intake, and distributions of cases and person-years (when available), and RRs or HRs with 95% CIs for ≥3 exposure categories were included. The median or mean intake of coffee in each category was assigned to the corresponding RR or HR with the 95% CI for each study. If coffee consumption was reported in a range of intake, we used the midpoint of the range. Similarly, when the highest category was open-ended, we assumed that the width of the category would be the same as the adjacent category. In the case of an open-ended lowest category, we clearly set the lower boundary to zero. Two-stage random-effects dose–response meta-analysis was performed to examine linear and non-linear relationship between coffee intake and risk of endometrial cancer. In the first stage, the method of Greenland and Longnecker  was used to calculate study-specific coefficients on the basis of results across categories of coffee intake, taking into account the correlation within each set of retrieved RRs and HRs ,24. ThroThe search identified 598 studies, of which 511 were excluded after reviewing the title, and 70 on the basis of the abstract . Of the I2 = 28%, p = 0.19, .The summary Relative Risk (RR) of endometrial cancer for the highest versus the lowest category of coffee consumption was RR = 0.79, 95% CI: 0.73, 0.87, with small heterogeneity  = 0.19, ; no publ = 0.19, . When weThe associations for caffeinated and decaffeinated coffee were consistent with the aforementioned relationship  was found for postmenopausal women, but not for premenopausal women . No differences were observed between ever-smokers and non-smokers, i.e., in both subgroups the risk was significantly reduced for the highest category of coffee consumption compared to lowest category  women .When taking into account the menopausal status, a significant decrease in the risk of endometrial cancer .Various meta-analyses on the association of coffee consumption and endometrial cancer risk have been conducted so far: for instance, in 2011 Je and Giovannucci  documentCoffee constituents have been associated with several biological mechanisms related to carcinogenesis, both in vitro and in vivo. These mechanisms include: DNA methylation, oxidative damage, activation of proto-oncogenes and inactivation of onco-suppressor genes, loss of apoptosis and growth control, and induction of angiogenesis ,38. ActiWith particular referral to female reproductive cancers, several mechanisms have been proposed: for instance, caffeine and coffee intake have been positively associated to sex hormone-binding globulin (SHBG) in postmenopausal women . SHBG isAdditional effects of coffee and caffeine intake on hormonal functions have been seen in improved insulin sensitivity as a result of the stimulation of insulin-mediated uptake of glucose . Coffee As already pointed out, we observed a stronger association between coffee consumption and endometrial cancer in women with high BMI (above 30) compared with women with a BMI of 25 or lower. Overweight and obesity have been associated with the development of cancers , and varThe present meta-analysis has some limitations. First, despite we provided insights on underrated factors potentially affecting the association between coffee consumption and endometrial cancer risk, other variables  remained largely unexplored. Second, we cannot rule out the possibility of changes in dietary habits  over time, leading to the risk of reverse causation in the event that an individual changed coffee intake due to a newly diagnosed medical condition.In conclusions, our findings suggest that increased coffee consumption is associated with decreased risk of endometrial cancer, especially in postmenopausal, obese women.


Association between the MGP gene rs1800801, rs1800802, rs4236 polymorphisms and vascular calcification and atherosclerotic disease was inconsistent. To clarify precise association, we performed this meta-analysis. Medline, Embase and China Knowledge Resource Integrated Database were systematically searched through December 2016. A total of 23 case-control studies, consisting of 5280 cases and 5773 controls, were included. The overall results suggested that the -7A polymorphism was associated with an increased risk for vascular calcification and atherosclerotic disease in the recessive model . Subgroup analyses of Caucasians showed significant associations in the allelic model, recessive model, and homozygote model: allelic model , recessive model , homozygote model . Subgroup analysis of the Asian population did not demonstrate any significant associations in any of the genetic models. No significant association was found in any genetic model amongst the rs1800802 and rs4236 polymorphisms. The findings of this meta-analysis indicate that the MGP gene rs1800801 polymorphism is significantly associated with vascular calcification and atherosclerotic disease, especially in the Caucasian population. Vascular calcification occurs as a part of the atherosclerotic process and it is an active process regulated similarly to the process of bone formation4. According to the current theoretical knowledge, the formation of microcalcifications causes plaque instability and is correlated with cardiovascular risk5. Therefore, vascular calcification is supposed to be a strong predictor of cardiovascular events independent of the traditional risk factors7.Atherosclerotic disease includes coronary artery diseases, cerebrovascular disease, and peripheral arterial diseases. It is still the leading cause of morbidity and mortality worldwide9. The mechanism by which MGP inhibits vascular calcification is still unknown. According to the current understanding, MGP regulates vascular calcification by binding and inactivating bone morphogenic protein 2 and preventing the deposition of calcium phosphate in the vascular matrix12.Matrix γ-carboxyglutamic acid Gla protein (MGP), a 10-kDa vitamin K-dependent extracellular matrix protein, has been shown to be an inhibitor of vascular calcification9. Genes rs1800801 (G7-A), rs1800802 (T138-C), rs4236 (Thr83-Ala) were most often reported. The association between MGP gene rs1800801, rs1800802, rs4236 polymorphisms and vascular calcification and atherosclerotic disease has been discussed in several studies, but the results have been controversial17. These results were inconclusive and did not reach a consensus. Therefore, we conducted this meta-analysis in order to precisely elucidate the genetic roles for the MGP gene rs1800801, rs1800802, rs4236 polymorphisms in the process of vascular calcification and atherosclerotic disease.Over the past decade, increasing evidence has indicated that several single nucleotide polymorphisms (SNPs) of the MGP gene may play a crucial role in the susceptibility of vascular calcification and atherosclerotic disease18. Relevant articles were identified by a systematic search of Medline, Embase and China Knowledge Resource Integrated (CNKI) Database from their inception to December 2016. The following search terms were used: “Matrix Gla protein”, “MGP”, “NTI”, “GIG36”, “MGLAP”, “polymorphism”, “polymorphisms”, “calcification”, “atherosclerosis”, “acute coronary syndrome”, “myocardial infarction”, “stenosis”, “ischemic stroke”, and “cerebral infarction”. Two authors independently confirmed the eligibility of articles and collated the data from the qualifying articles. The reference lists of retrieved articles were also reviewed for eligible studies. There were no language restrictions.This meta-analysis was performed according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) criteriaEligible articles should meet the following criteria: (1) evaluated the association of the MGP gene  polymorphism with vascular calcification and atherosclerotic disease, (2) studied on human beings, (3) in a case-control or nested case-control study design.We excluded studies according to the following criteria: (1) unrelated to the association of MGP gene polymorphism with vascular calcification and atherosclerotic disease; (2) review articles; (3) have no control group; (4) animal studies; (5) data is missing or incomplete, and the authors could not be contacted; (6) data is duplicated.19. Any study with a score greater than 7 was considered as “high quality”. Disagreements were resolved by consensus or arbitration by a third reviewer.Relevant information was carefully extracted from all eligible articles. The following data were extracted: first author, year of publication, country of origin, ethnicity, source of controls, frequency of genotypes in cases and controls, and evidence of Hardy-Weinberg equilibrium (HWE) in controls. Two authors independently extracted the data and assessed the study quality based on the Newcastle Ottawa Scale (NOS)20. Odds ratios (ORs) and 95% confidence intervals (CIs) were used to assess the strength of associations of the MGP gene  polymorphism with vascular calcification and atherosclerotic disease. The Z test was used to assess the significance of the ORs, and a P value < 0.05 was considered statistically significant. Heterogeneity between studies was tested through chi-square and I-square (I2) tests. A fixed-effects model was used if the I2 value was less than 50% and the p-value was greater than 0.1; otherwise a random-effects model was used. Subgroup analyses were conducted based on ethnicity and source of control. Sensitivity analyses were performed to display possible variability. Begg’s and Egger’s linear regression tests were applied to assess the potential publication bias22.All analyses were computed in Stata software version 12 . HWE was assessed for each SNP among controls using a χ2 test27 met the inclusion criteria, including 23 studies consisting of 5280 cases and 5773 controls. Among these included studies, 18 were performed in the Caucasian population and 5 were performed in Asians. All were case-control studies and had been published between 2000 and 2016. The characteristics of eligible studies are shown in Table The study selection process is shown in Fig. Distribution and allele frequency of the three MGP gene polymorphisms in the cases and controls are shown in Table Nine studies with a total of 2015 cases and 2423 controls evaluated the association of the rs1800801 polymorphism with vascular calcification and atherosclerotic disease. There was a significant association of rs1800801 gene polymorphism with vascular calcification and atherosclerotic disease in the recessive model  , dominant model , recessive model , homozygote model , and heterozygote model  , dominant model , recessive model , homozygote model , and heterozygote model . Similar findings were seen in the subgroup analyses in different ethnicities and source of controls , dominant model , recessive model , homozygote model , and heterozygote model . The subgroup analyses in different ethnicities and source of controls showed similar results  is located on the short arm of chromosome 12 (12p12.3). There is increasing evidence that genetic variation at the MGP locus could modulate the development of vascular calcification and atherosclerotic disease. Previous studies have shown that MGP genes rs1800801, rs1800802 and rs4236 polymorphisms have an important impact on the promoter activityIn this meta-analysis, we investigated the association between three SNPs in the MGP gene with the risk of vascular calcification and atherosclerotic disease in 23 case-control studies (consisting of 5280 cases and 5773 controls). The overall results revealed that only the rs1800801 polymorphism was associated with the risk of vascular calcification and atherosclerotic disease. Stratification analysis by ethnicity indicated that the association was significant among Caucasians, but not among Asians in rs1800801 polymorphism. The reason why this association varies among different ethnicities is not clear, the small number of studies or the natural selection in different ethnicities may explain it. No significant association was found in the rs1800802 and rs4236 polymorphisms. Stratification analyses by ethnicity and source of control showed similar results in the rs1800802 and rs4236 polymorphisms.in vitro study revealed that the rs1800801–7A variant had an approximately 1.5-fold higher activity than -7G variant in VSMCs30. The -7A allele occurred more frequently in patients with vascular calcification, myocardial infarction, and ischemic atherothrombotic stroke25. Therefore, the -7A allele of the MGP gene may confer an increased risk of vascular calcification and atherosclerotic disease, and therefore may be a novel promising target for prevention and treatment.The 31. While another study found that the rs1800801 polymorphism was associated with an increased risk of ischemic atherothrombotic stroke only in women25. However, since detailed gender specific data could not be obtained for the studies included in this meta-analysis, we were unable to perform a sub-analysis by gender. Second, as the heterogeneity in different ethnicities influenced the results significantly, the stratification analysis of the source of control was not conducted in the SNP rs1800801. Third, only three studies were performed in Asians, therefore the findings from the Asian based studies were not convictive enough; more studies focusing on the Asian population are needed.There are several limitations to this study. First, the association of the MGP gene polymorphism with vascular calcification and atherosclerotic disease may be influenced by gender. In some studies, the association of MGP polymorphisms with vascular calcification and atherosclerotic disease was only observed in menIn conclusion, findings from this meta-analysis indicate that the MGP rs1800801 polymorphism is associated with an increased risk of vascular calcification and atherosclerotic disease. Furthermore, this association might only exist in Caucasians. The MGP gene rs1800802 and rs4236 polymorphisms are not associated with an increased risk of calcification and atherosclerotic disease. A larger number of epidemiological studies are required to confirm our findings.


C-reactive protein (CRP) gene rs1205 polymorphism and the risk of colorectal cancer (CRC) has been investigated previously. However, the results were conflicting. In the present study, we assessed whether CRP gene rs1205 polymorphism was associated with the risk of CRC by meta-analysis. We searched in PubMed, Embase, and the CNKI databases. Pooled odds ratios (ORs) and 95% confidence intervals (CIs) were calculated. Seven original studies involving 4,181 cases and 10,601 controls analyzed the association between CRP gene rs1205 polymorphism and CRC risk. No significant association was found between CRP gene rs1205 polymorphism and CRC risk in this meta-analysis. Sensitivity analysis did not draw different findings. Stratification analyses of ethnicity, type of cancer, and genotype method also did not obtain any association between CRP gene rs1205 polymorphism and CRC risk. In conclusion, this meta-analysis indicates that CRP gene rs1205 polymorphism was not associated with the risk of CRC.The relationship between  Colorectal cancer (CRC) is the third most common cancer and the fourth most common cancer cause of death globally . The patCRP gene is located at chromosome 1q21–1q23 consisting of two exons and spans 1.9 kb in length, including 29 single nucleotide polymorphisms (SNPs). CRP is associated with a wide range of diseases, including atherosclerosis and diabetes mellitus [CRP may be a candidate gene for CRC susceptibility.C-reactive protein (CRP) is the phenotype acute-phase protein induced by hepatocytes, known as an inflammatory biomarker. The mellitus ,8. Givenmellitus ,9–14. ReCRP gene rs1205 polymorphism and CRC risk [CRP gene rs1205 polymorphism in the development of CRC, we performed a comprehensive meta-analysis to clarify the association between this SNP and CRC risk.Recently, a lot of studies explored the relationship between CRC risk –20. HoweCRP gene rs1205 polymorphism and CRC risk; (2) studied on human beings; (3) contained genotype data for the calculation of odds ratios (ORs) and 95% confidence intervals (CIs). Related information was carefully extracted from all eligible studies. The following information was extracted from each study: author, year of publication, ethnicity based on the continent of origin of the study population, source of controls (SOC), numbers of cases and controls, and the genotype methods.We performed a comprehensive literature search throughout PubMed, Embase, and CNKI databases to retrieve the genetic association studies of CRC. The following terms were used in our searching strategies: ‘c-reactive protein’, ‘CRP’, ‘SNP’, ‘polymorphism’, ‘variant’, ‘cancer’, ‘carcinoma’, and ‘malignancy’. Additional potential omitted studies (such as reference lists of identified studies) have been identified by hand screening. All studies were carefully selected and were up to date as of May 1, 2017. The inclusion criteria for studies were as follows: (1) studies that evaluated the association between CRP gene rs1205 polymorphism and CRC risk. Stratification analyses were carried out by SOC. P<0.05 was considered statistically significant. Multivariate ORs and corresponding 95% CIs between extreme levels of annualized case volume (highest versus lowest) were pooled using a random-effects model, accounting for clinical heterogeneity. Heterogeneity across studies was assessed by using the Q statistic with its P value and I2 statistic [All statistical analyses were performed using the Stata 11.0 software . ORs and 95% CIs were used to assess the strength of associations between tatistic ,22. Pooltatistic . Two revtatistic . All disP<0.05 was considered to indicate statistically significant. We performed sensitivity analysis by omitting each study in turn to determine the effect on the test of heterogeneity and evaluated the stability of the overall results.Potential publication bias was assessed by Begg’s and Egger’s linear regression test . P<0.05 As showed in CRP gene rs1205 was not associated with CRC risk , P=0.421; TT versus CC: OR and 95% CI, 1.14 , P=0.257; TT + CT versus CC: OR and 95% CI, 1.03 , P=0.758; TT versus CT+CC: OR and 95%CI, 1.16 , P=0.078; TC versus CC: OR and 95% CI, 0.99 , P=0.866, In the general analysis, we found that CRP rs1205 polymorphism (We assessed sensitivity analysis by omitting each study one at a time in every genetic model for rs1205 polymorphism. The pooled ORs for the effects of the SNP on the risk for CRC risk indicated that our data were stable and trustworthy. Both Egger’s and Begg’s tests were used to evaluated the publication bias of this meta-analysis. Our data revealed that there was no obvious publication bias for morphism .CRP gene rs1205 polymorphism and CRC susceptibility. CRP is one of the most common acute-phase proteins induced by hepatocytes. Plasma CRP level may dramatically increase by up to 10,000-fold at the time of acute responses to severe tissue damage or serious infection [CRP gene rs1205 polymorphism and risk of CRC, but the results were inconsistent [CRP gene rs1205 polymorphism was not associated with CRC risk. Stratification analyses of ethnicity, type of cancer, and genotype method also did not obtain any association between this SNP and CRC risk.To our best knowledge, this is the first quantitative assessment of the genetic association studies reporting on the relationship between nfection . Severalnsistent –20. ThisCRP rs1205 polymorphism and CRC risk. However, Siemes et al. [CRP rs1205 polymorphism with CRC risk and different ethnicities. Two meta-analyses [CRP gene rs1205 polymorphism with cancer susceptibility previously. Zhang et al. [CRP rs1205 polymorphism and the risk of overall cancer. However, in subgroup analysis by cancer type, marginally increased risk was observed in CRC. Geng et al. [CRP gene rs1205 polymorphism and CRC risk. Second, we identified seven studies [A single study could be underpowered because of sample size, diversity inheritance of the heterogeneous, different ethnicities, clinical heterogeneity, and so on. For instance, Nimptsch et al. , Ognjanos et al.  failed tanalyses ,28 invesg et al.  found nog et al.  did not g et al. ,20, whilg et al. ,17,19,20 studies –20 with CRP gene rs1205 polymorphism and the clinical manifestations of CRC.Several potential limitations should be addressed in this meta-analysis. First, the heterogeneity of this meta-analysis is high, so the data should be interpreted with caution. Second, due to limited data, we could not conduct further stratification analyses of other potential factors, such as age, gender, smoking, and alcohol consumption. Third, our results were based on unadjusted estimates for confounding factors, which might have affected the final results. Fourth, we could not assess potential gene–gene and gene–environment interactions because of the lack of relevant data. Fifth, the conclusions of some stratification analyses about rs1205 polymorphism should be interpreted with caution due to limited sample size. Sixth, the sample sizes of some stratification analyses were limited. Finally, we cannot examine the association between CRP gene rs1205 polymorphism is not associated with the risk of CRC. Further studies with large sample size is necessary to validate whether CRP gene rs1205 polymorphism contribute to CRC susceptibility.In conclusion, this meta-analysis confirms that


Lynch syndrome (LS) is a highly penetrant inherited cancer predisposition syndrome accounting for approximately 1000 cases of colorectal cancer (CRC) in the UK annually. LS is characterised by autosomal dominant inheritance and germline mutations in DNA mismatch repair genes. The penetrance is highly variable and the reasons for this have not been fully elucidated. This study investigates whether low penetrance genetic risk factors may result in phenotype modification in LS patients. To conduct a systematic literature review and meta-analysis to assess the association between low penetrance genetic risk modifiers and CRC in LS patients. A systematic review was conducted of the PubMed and HuGENet databases. Eligibility of studies was determined by pre-defined criteria. Included studies were analysed via the per-allele model and assessed by pooled odds ratios and establishing 95% confidence intervals. Study heterogeneity was assessed via Cochrane’s Q statistic and I2 values. Publication bias was evaluated with funnel plots. Subgroup analysis was conducted on gender. Statistical software used was the Metafor package for the R programme version 3.1.3. Sixty-four polymorphisms were identified and sufficient data was available for analysis of ten polymorphisms, with between 279 and 1768 CRC cases per polymorphism. None demonstrated association with CRC risk in LS patients. However in sub-group analysis the polymorphism rs16892766 (8q23.3) was significant in males . The variable phenotype presentation of the disease still remains largely unexplained, and further investigation is warranted. Other factors may also be influencing the high variability of the disease, such as environmental factors, copy number variants and epigenetic alterations. Investigation into these areas is needed as well as larger and more definitive studies of the polymorphisms analysed in this study. Colorectal cancer (CRC) development is highly complex and multi factorial involving lifestyle, environmental and genetic factors. The concordance rates of cancer in monozygous and dizygous twins suggest that about 35% of the variation in cancer risk might typically be ascribed to heritable factors . Highly MLH1 and MSH2 which together account for 70–80% of cases, MSH6 which is approximately 20% and PMS2 which is associated with lower penetrance  polymorphism” AND “Lynch syndrome” OR “HNPCC” OR “Hereditary Nonpolyposis Colorectal Cancer”.An additional search of the HuGENet database was carried out (last search August 2016) using the terms “Gene name” AND “Lynch syndrome”.Where a gene has two commonly used names both were used during the literature search. The results obtained from the second search were integrated with the first search results.Further relevant articles were studied from the bibliographies of eligible articles and previously conducted meta-analysis that looked at polymorphisms related to CRC to identify any further relevant studies to be included.An example by which studies were included through screening and determining eligibility is shown in Fig. Whether or not the study was deemed eligible for inclusion was based upon pre-defined inclusion and exclusion criteria and by mutual agreement. Studies had to meet ALL the inclusion criteria and were excluded if they met ANY of the exclusion criteria as show in Table For included studies, data extraction and quality appraisal was performed by one reviewer (ND) and checked by a second (KM). Any disagreements were resolved by discussion, with involvement of a third reviewer as necessary.Once a study was deemed eligible for inclusion the following data was acquired using a standardised database: Name of lead author, publication year, paper title, polymorphism or polymorphisms investigated, minor allele frequency, number of cases, number of controls, genotype frequencies for cases, genotype frequencies for controls, genotype frequencies for MMR mutation subtypes for cases and controls, genotype frequencies for male and female cases and controls, diagnosis method of cases and controls, country of study origin and ethnicity of study participants.Statistical analysis was undertaken using the Metafor package in R (version 3.1.3) .Hardy–Weinberg equilibrium (HWE) was assessed using the Pearson’s Chi-squared test which tested for variation between the expected genotype frequencies and the observed frequencies. A value above 3.84 (p < 0.05) was regarded as being statistically significant for variation between the two frequencies. All included studies were calculated to be in HWE.Quantitative synthesis was achieved by investigating pooled odds ratios (OR), and 95% confidence intervals (CI). The allele specific model was utilized in OR calculations. Polymorphisms were found to be statistically significant if the 95% CI did not cross 1.Study heterogeneity was measured with Cochrane’s Q statistic and I² values. Guidelines suggest that 25, 50 and 75% indicate low, medium and high heterogeneity respectively .If I² values were 50–100 indicating high heterogeneity the DerSimonian and Laird random effects method is suggested , 20. WheSensitivity analysis was conducted on suitable polymorphisms (>3 studies).Publication bias was assessed using funnel plot asymmetry . Egger’sSubgroup-analyses were conducted if there were greater than 3 studies per subgroup of gender, ethnicity or MMR mutation subtype. Sub-group analyses were possible for gender for two single nucleotide polymorphisms (SNPs).Sixty four polymorphisms in 39 genes were eligible for inclusion. Insufficient data for meta-analysis was available for 53 polymorphisms. This was due to a small number of studies being present for each SNP, with only one study available for 45 polymorphisms and two available for eight polymorphisms. Polymorphisms which could not be analysed are not presented in this paper but are available upon request.CYP17A1 (rs743572), KIF20A (rs10038448), CDC25C (rs6874130), KDM3B/FAM53C (rs3734168) and SMAD7 (rs4939827).Out of the 53 excluded polymorphisms 48 showed no significant difference, while five demonstrated a significant change in ORs. These were for GSTT1 (null variant) displayed a low level of heterogeneity (I² = 13.92) and CYP1A1 (rs4646903) displayed a high level of heterogeneity (I² = 77.07).For most polymorphisms, studies were homogenous. However two polymorphisms displayed heterogeneity—It was possible to conduct meta-analyses on ten polymorphisms in five genes and five un-associated SNPs. Studies which were eligible and were included in the meta-analysis are shown in Table Forest plots were created to investigate the pooled ORs, 95% CI and individual study weights. We did not find any polymorphism that was statistically significant for any change in the risk of CRC development.Insufficient data was reported for subgroup analysis regarding ethnicity. However adequate data was received in order to conduct gender subgroup analysis on two polymorphisms and MLH1 gene mutation subtypes.Gender subgroup analysis was possible for two SNPs. These were for rs16892766 and rs3802842, and results are presented in Table The overall OR for rs16892766 is not significant . However when subgroup analysis was conducted on the individual genders the OR for males moved into being statistically significant. When a comparison was made between males and females positive for a diagnosis of CRC it was indicated that males with this polymorphism will have a higher risk of developing CRC than females. It could be inferred from this data that the minor allele (C) increases the risk of males developing CRC and that the risk allele for this polymorphism is C. No significant association was found with the polymorphism rs3802842 for either males or females.MLH1 with values for rs16892766 being OR 1.12, 95% CI 0.79–1.59, values for rs3802842 being OR 1.22, 95% CI 0.97–1.53.Analysis was possible for the SNPs rs16892766 and rs3802842. This was possible through extraction of data from Talseth-Palmer et al.  which inP53 (rs1042522), CCND1 (rs9344), GSTM1, rs16892766 and rs3802842. Each study was removed in succession and ORs and 95% CI values were recalculated. No significant changes in pooled ORs, 95% CI or study heterogeneity were observed.Sensitivity analysis was possible for five polymorphisms; CCND1 is shown in Fig. CYP1A1 was an anomaly and was asymmetrical. Due to the low number of studies included in the analysis the study bias is difficult to predict.Nine of the polymorphisms studied did not show publication bias and showed symmetry in a funnel plot. The funnel plot for The aim of this study was to investigate the association between genetic risk modifiers and the risk of developing CRC in a LS patient cohort. The results indicate that polymorphism rs16892766 is statistically significant in males, with the minor allele (C) increasing the risk of development of CRC. The rest of the polymorphisms investigated did not achieve significance, suggesting that the role played by these risk modifiers is a minor one. However due to some polymorphisms being close to significance more study into this area and a larger data pool would help to provide increased certainty into any possible role including greater insight into specific MMR gene mutation subtypes to discern whether the minor role still holds true.P53 functions as a tumour suppressor gene regulating cell cycle control, apoptosis and DNA integrity . ResultsCCND1 gene encodes for Cyclin D1 which forms the regulatory subunit of CDK4/6 enzyme. This can cause tumourgenesis by directly phosphorylating the tumour suppressor protein retinoblastoma (Rb); this leads to cells being more likely to bypass the G1 to S phase cell cycle checkpoint which leads to cancer development [CCND1 polymorphism (rs9344) which is consistent with other literature sources. Zhang et al. conducted a meta-analysis on CCND1 and CRC susceptibility [The elopment . There whism rs93 which ishism rs93 which isMLH1 mutation carriers with this polymorphism do have an increased risk of developing CRC [MLH1 mutation carriers. There has been no evidence of MSH2 carriers having an altered risk due to rs16892766 or rs3802842 mutations. The other three polymorphisms  that were obtained via GWAS studies did not alter risk of CRC development.Rs16892766 overall did not reach significance; however results were very close to reaching significance, the forest plot of this is shown in Fig. ping CRC . Overallping CRC . Our anaGSTM1 and GSTT1 are two subtypes of the Glutathione-S-transferase family of enzymes that can protect against developing cancer [GSTM1 the null variant is the most common polymorphism, which results in reduced enzymatic activity and has been associated with the development of cancers [g cancer . They arg cancer , 47. In  cancers . Previou cancers , 48. ThiGSTT1 we did not find an association between risk of CRC development and the null variant. There was a small amount of study heterogeneity (I² = 13.92) which was classified as low study heterogeneity. It was not appropriate to remove any studies increasing heterogeneity due to only three studies being present. Results were similar to what was found for GSTM1, where some meta-analyses have found an association with very low 95% CI [For the gene w 95% CI , 49. FurCYP1A1 and CRC development. As only three studies were included it was not possible to conduct sensitivity analysis on the data to address the high heterogeneity. Zheng et al. concluded that this polymorphism of CYP1A1 is a low penetrance modifier of CRC development [No association was found between elopment . Due to EIF3H [EIF3H gene codes for the H subunit of eukaryotic translation initiation factor 3 (EIF3). It is thought to be involved in protein synthesis and overexpression results in increased proliferation, growth and cell survival [Subgroup analysis for gender was possible for two of the polymorphism . The rs16892766 polymorphism when analysed for males was significant. However for females there remained no effect. When a direct comparison was made between males and females, results were significant for an increased OR for males. The polymorphism has not been mapped to any particular gene, although it has been mapped to a linkage disequilibrium block that contains the gene EIF3H . The EIFsurvival .Candidate polymorphisms were often not included in the meta-analysis due to a small number of studies investigating their effect in a LS patient population. They were still found in the systematic review and data was extracted from them and analysed. Four studies were found with statistically significant results and are shown in Table CYP17A1 polymorphism (rs743572) and found it to increase risk of development of CRC [MSH2 were included in the study. CYP17A1 belongs to the cytochrome p450 family and is involved in the metabolism of endogenous compounds and sex hormones and has been associated with cancers [Campbell et al. investigated a t of CRC . However cancers .DM3B/FAM53C (rs3734168) and CDC25C (rs6874130) to be significantly associated with the development of CRC [CDC25C (rs3734166) and the possibility of being correlated with it. Win et al. observed a statistically significant association between a polymorphism in SMAD7 (rs4939827) and CRC development [SMAD7 has effects on the TGF-b pathway. TGF-b acts as a tumour suppressor in early states of tumour formation [SMAD7 in CRC development for LS should be investigated further due to conflicting data.Chen et al. found polymorphisms in t of CRC . Howeverelopment . SMAD7 hormation . Anotherormation . The effA large number of candidate polymorphisms could not be included due to a lack of sufficient eligible number of studies. For the polymorphisms that did qualify for inclusion, there were still only a relatively small number of studies.Some studies had a small number of participants and were excluded for not meeting the inclusion criteria.Another difficulty encountered was that data was frequently missing from studies. On occasion these were the genotype frequencies, meaning that some studies did not meet the inclusion/exclusion criteria on these grounds. More commonly data was missing in the form of MMR mutation subtypes and information regarding ethnicity of study participants.Overall this study did not identify any statistically significant association in ten polymorphisms for the development of CRC in LS patients. However, upon conducting gender specific sub group analysis one polymorphism (rs16892766) at chromosome locus 8q23.3 was significant for increasing the risk of males with the minor allele C. For this polymorphism it was found that C is the risk allele. The variable phenotype presentation of the disease still remains largely unexplained, and further investigation is warranted. Other factors could also be influencing the high variability of the disease, with environmental factors, copy number variants and epigenetic changes possibly having an influence, and investigation into these areas is also needed. Another factor we were not able to fully study were gene–gene interactions whereby patients with multiple low penetrance SNPs could be experiencing an additive effect to increase risk. However we conclude that there is currently no consistent evidence that the phenotype of Lynch syndrome is influenced by the effects of low penetrance modifiers.


Severe fever with thrombocytopenia syndrome (SFTS) is an emerging infectious disease caused by a novel bunyavirus-SFTSV. The seroprevalence of anti-SFTSV antibodies including immunoglobulin G (IgG) and immunoglobulin M (IgM), specific to SFTSV in the general population has been investigated in various epidemiological studies with inconsistent results. Here, we clarify this discrepancy and reach a more comprehensive result by mean of a meta-analysis.All relevant articles were searched in the electronic databases  up to November 2016. The pooled seroprevalence and 95% confidence intervals (95% CIs) were calculated by random- or fixed- model on the basis of heterogeneity.In total, 21 studies containing 23,848 blood samples from 7 provinces were included in this meta-analysis. The minimum and maximum reported seroprevalences of SFTSV among humans in China were 0.23% and 9.17%, respectively. The overall pooled seroprevalence of SFTSV antibodies was 4.3% (95%CI: 3.2%-5.5%). The pooled prevalence was 5.9% (95%CI: 4.7%-7.0%) in Zhejiang province, 4.9% (95%CI: 4.1–5.8%) in Anhui province, 3.9% (95%CI: 1.3%-6.4%) in Shandong province, and 0.7% (95%CI: 0.2%-1.1%) in Jiangsu province. Stratified by occupation, the pooled prevalence of farmer was 6.1% (95%CI: 3.4%-8.9%) and others (mainly are students) was 3.3% (95%CI: 2.4%-4.2%). Additionally, seroprevalence of SFTSV in people who lived in the same village with the patient were higher than that of people who lived in a different village. Seropositive rates in sampling years after 2012 were higher than that before 2012. The prevalence of SFTSV did not differ by age or gender. Sensitive analysis by omitting one study at a time indicated the results of the pooled seroprevalence were robust.Seroprevalence of SFTSV among healthy population in central and eastern China is high. Surveillance efforts on mild or asymptomatic infections among endemic persons are needed. Severe fever with thrombocytopenia syndrome (SFTS) is a notifiable infectious disease characterized by fever, weakness, leukopenia, thrombocytopenia, gastrointestinal symptoms, and central nervous system manifestations –3. The cTime range from March to November is the epidemic season of SFTS, May-July is the peak time of SFTSV infection. SFTSV is most likely to be transmitted by tick bite according to evidence from tick exposure history and SFTSV detection in ticks such as Haemaphysalis longicornis ticks, and virus gene sequence analysis showed the virus in ticks closely related to those circulating in humans –10. HighInfluenza A (H9N2), Ebola and Marburg viruses, and Enterovirus 71, etc [In past years, seroprevalence of SFTSV among healthy population has been widely investigated in various epidemiologic studies. Current estimates of the seroprevalence of SFTSV among general humans in China are almost based on one or several villages with relatively small sample size in one province rather than nationally representative sample of this population. In addition, seroprevalence of SFTSV among healthy humans in different gender, age groups, and endemic regions has yet to be illuminated. Meta-analysis is an effective method of pooling data of individual study together, thus enhancing the statistical power of the analysis for the estimates, to reach a more comprehensive result. It has been widely utilized in sero-epidemiological studies of infectious diseases, such as  71, etc –18. In pAll relevant articles about the seroprevalence of SFTSV in general population were searched via the following electronic databases: Pubmed, Web of Science, Embase, Chinese National Knowledge Infrastructure (CNKI), Chinese WanFang Database. Date searches were carried out up to November 2016, without restrictions regarding language, publication year and district. The following search words and terms in English and Chinese were used in the final search:  and (“SFTS” or “SFTSV” or “Severe fever with thrombocytopenia syndrome” or “Severe fever with thrombocytopenia syndrome virus”). Moreover, all relevant references cited in original articles and reviews were also manually searched to identify additional articles not indexed by these databases.To obtain the valid articles we needed, the following criteria were established: 1) the study either reported the seroprevalence of SFTSV or had sufficient data for calculating the seroprevalence; 2) individuals neither had SFTS infection in the past, being hospitalized for any clinically similar disease, or contact with a person who had SFTS as defined previously [After initial evaluation, two authors independently and carefully screened the articles on title and abstract according to the eligibility criteria. Authors also filled out a standard quality assessment checklist with 11 items concerning the methodological aspects of cross-sectional studies for each study . If the The quality of the included studies was assessed by Agency for Healthcare Research and Quality (AHRQ), which is consisted of 11-item with a yes/no/unclear response option: “No” or “unclear” was scored “0” and the “Yes” would be scored “1”. ArticleAccording to the results of heterogeneity test in different groups, the proper model was adopted to assess a pooled value and 95%CI of seroprevalence of SFTSV antibodies in healthy population.2 statistics [P-values of the Cochran Q test less than 0.1 is considered to be statistically significant, and I2 value more than 75% indicates high heterogeneity. A significant Q statistic  indicate heterogeneity across studies, and then the DerSimonian and Laird method in random effect model result is used for meta-analysis. When the P-value of heterogeneity test is more than 0.1 , the results of Mante-Haenszel method in a fixed-effect model will be adopt. In order to explore the potential source of heterogeneity between studies and the seroprevalences of SFTSV antibodies with different characters such as gender, age, occupation and province, subgroups analysis were also conducted. Additionally, a leave-one-out sensitivity analysis was carried out to assess the impact of each study on the overall pooled estimate.Statistical heterogeneity across different studies was assessed by using the Cochran Q and Iatistics . P-valueP-value of Egger’s test was less than 0.05, statistically significant publication bias might exist.The potential publication bias was assessed by using Begg’s funnel plot . Funnel All statistical analysis were performed using STATA 11.0 .Table 1.The detailed study retrieval steps according to the PRISMA statement was shown in Table 1). Generally, the most prevalence studies were conducted in the center and east of China , the, theTablFig 3. The heterogeneity across the studies was high . The overall pooled seroprevalence estimate for SFTSV in random effect model was 4.3% (95%CI: 3.2%-5.5%).Seroprevalence of SFTSV antibodies varied from 0.23% to 9.17% and were displayed as forest plots in Table 2. Stratification by gender, the pooled prevalence of male was 5.3% (95%CI: 3.3%-7.2%), and female was 5.2% (95%CI: 3.4%-7.0%). Stratification by age, the pooled seroprevalence of less than age of 40 years was 4.1% (95%CI: 2.4%-5.7%), and more than age of 40 years was 4.9% (95%CI: 2.9%-7.0%). Stratification by occupation, the pooled seroprevalence of farmer was 6.1% (95%CI: 3.4%-8.9%) and others (mainly are students) was 3.3% (95%CI: 2.4%-4.2%). Seropositive rates were comparable among sampling years, seroprevalence in the years before 2012 and after 2012 was 3.1% (1.9%-4.4%) and 5.2% (4.6%-5.9%), respectively. Stratification by testing method, seroprevalence in the method of I-ELISA and D-ELISA was 6.2% (5.4%~7.1%) and 1.8% (1.1%~2.5%), respectively. Seroprevalence of SFTSV in people who lived in the same village with the patient and lived in a different village was 6.5% (4.7~8.3) and 4.1% (3.1%~5.1%), respectively. We also found seroprevalence in endemic area was 4.3% (3.1%~5.6%) and in rural village was 4.3% (3.2%~5.4%). Additionally, the pooled seroprevalence was 5.9% (95%CI: 4.7%-7.0%) in Zhejiang province, 3.9% (95%CI: 1.3%-6.4%) in Shandong province, 4.9% (95%CI: 4.1–5.8%) in Anhui province, 6.9% (5.5%~8.2%) in Hubei province and 0.7% (95%CI: 0.2%-1.1%) in Jiangsu province. Forest plots for the meta-analysis in different provinces were shown in Fig 4.To explore the potential source of the high heterogeneity, we did some subgroup analysis by gender, age, occupation and province. The detail results of subgroup analysis were shown in Fig 5). The Egger’s test results demonstrated evidence of publication bias, however, the Begg’s bias test was not significant for studies reporting seroprevalence of SFTSV (data not shown). A leave-one-out analysis which was performed to assess the impact of the individual study on the pooled estimates. Sensitive analysis showed that no single study qualitatively altered the pooled seroprevalence estimates, providing evidence of the stability of the meta-analysis .Potential publication bias of the included articles were evaluated by Begg’s funnel plots and Egger’s test. The shapes of the funnel plot was asymmetry evidently which gives rise to suspected publication bias . It was higher than a recent investigation in Korean population with a seroprevalence of 2.1% based on 1069 serum samples . SensitiPrevious study reported gender ratio of SFTS patients in China was about 1:1.15 , but it varied among different provinces . In thisAge composition of SFTS patients was widely, previous studies had demonstrated that SFTS patients ranged from 1 to 93 years old, but most of patients were aggregated in 40–79 years . GeneralThe infections rates of healthy population varied in different provinces. It reflects that incidences of the latent infection of SFTS are different in endemic regions. In this study, we found SFTSV seroprevalence was high in Henan, Hubei, Zhejiang, Shanxi, Anhui and Shandong provinces, and was relatively low in Jiangsu provinces with seroprevalence of 0.7% (95%CI: 0.2~1.1). However, this result should be interpreted with cautious because of limiting number of study and sample size in some provinces, which may lead to lack of representativeness.Additionally, different ELISA assay such as indirect ELISA (I-ELISA) and double-antigen sandwich ELISA system (D-ELISA) used to detect SFTSV-specific antibodies might also contribute to heterogeneity. Compared to the traditional I-ELISA, D-ELISA was considered having more higher sensitivity to detect total antibodies . HoweverSeroprevalence was also observed between different years, seroprevalence in the year after 2012 was higher than that before 2012, indicating ongoing and intensified circulation of SFTSV in endemic areas of China. Another discrepancy may be attributed to the sampling season, sampling method, previous contacting with animals , or antibody types  detection in different studies.The advantage of this systematic review and meta-analysis is that good quality studies from many centers were pooled for a relatively large sample size, but there are several limitations also should be addressed. Firstly, significant heterogeneity was detected across studies. Although we did some subgroup analysis to identify sources of heterogeneity, many unmeasured factors may have influenced the results. Secondly, funnel plots and Begg’s tests indicated that publication bias might be exist in the present study, which may distort the estimates of seroprevalence, so that the results should be interpreted with cautions.In summary, data from 21 published studies suggested that SFTS is circulated widely in China and could be a cause of considerable health problems in the country. Surveillance efforts on mild or asymptomatic infections among endemic persons are needed. Further subtle-designed studies are still needed to describe the exact epidemiology of the disease at a national level in other parts of China.S1 Dataset(XLS)Click here for additional data file.S1 Table(DOC)Click here for additional data file.S1 Text(DOC)Click here for additional data file.