Helicobacter pylori is an important carcinogenic factor in gastric cancer. Studies have shown that Helicobacter pylori infection is inversely associated with certain diseases such as esophageal cancer and whose infection appears to have a “protective effect.” At present, the relationship between Helicobacter pylori infection and esophageal cancer remains controversial. This study was designed to investigate the relationship between Helicobacter pylori infection and the risk of esophageal cancer in different regions and ethnicities.  Systematic search of the articles on the relationship between Helicobacter pylori infection and esophageal cancer from the database with the duration time up to December 2018. This systematic review was performed under the MOOSE guidelines.  This meta-analysis included 35 studies with 345,886 patients enrolled. There was no significant correlation between Helicobacter pylori infection and esophageal squamous cell carcinoma in the general population . However, a significant correlation was found in the Middle East . There was no significant difference in the prevalence of Helicobacter pylori between the case group and the control group in esophageal adenocarcinoma (8.87% vs. 9.67%). The pooled OR was 0.55 (95% CI: 0.43-0.70) or 0.23 (95% CI: 0.15-0.36). When grouped by match or not, the pooled OR of the nonmatching group and the matching group was 0.48/0.21 (95% CI: 0.36-0.65/95% CI: 0.13-0.36) and 0.73/0.71 (95% CI: 0.57-0.92/95% CI: 0.60-0.84), respectively.  In the general populations, no significant association was found between Helicobacter pylori infection and the risk of esophageal squamous cell carcinoma. However, lower risk was found in the Middle East. Helicobacter pylori infection may reduce the risk of esophageal adenocarcinoma, but such “protection effect” may be overestimated. Esophageal cancer ranks the eighth in the world's cancer incidence and the sixth in the global cancer death cause . There aHelicobacter pylori is a common bacterium in the upper digestive tract, which infects about half of the world . MarshalThe data of this meta-analysis were collected based on the Meta-analysis of Observational Studies in Epidemiology (MOOSE) statements  .All articles and abstracts published up to December 2018 were systematically searched in Embase and PubMed using MeSH terms and free words. Some of the database search MeSH terms are the following: “Esophageal Neoplasms”, “Esophageal squamous cell carcinoma”, “Adenocarcinoma of Esophagus”, and “Helicobacter pylori”. Comprehensive search terms are listed in the appendix. Besides, manual searches were conducted to ensure that all articles were related to our subject. Other sources are from related articles mentioned in the previous related meta-analysis –20. TherThe two authors (HQ Gao and LN Li) independently selected articles to be included and conducted critical assessments. Discrepancies were resolved by reaching a consensus with the senior author (JY Jing). Eligible studies were selected according to the following inclusion criteria: (1) they are studies on the relationship between Helicobacter pylori infection and esophageal cancer, (2) the incidence of esophageal cancer in the control and case groups can be extracted when exposed to or not exposed to Helicobacter pylori, (3) the odds ratio (OR) (case-control study) or relative risk (RR) (cohort study) and 95% confidence interval (95% CI) can be directly or indirectly calculated, and (4) the full text can be obtained. The exclusion criteria are (1) unobtainable full text, (2) the repeated study of the same population sample, and (3) the secondary data analysis literature such as review or meta-analysis.Data extraction was independently completed by the two researchers (HQ Gao and LN Li). First of all, after the thorough review of the title and abstract, obtain the full-text literature which meets the inclusion criteria. Two sides discuss for the controversial literature and seek the third senior researcher (JY Jing) to make a final decision if controversies were meet. The following data were extracted from each article: first author's name, year of publication, study country, ethnicity, study type, study subject, Helicobacter pylori test method, and number of control and case groups, as well as related measurement indicators (OR or RR and 95% CI).Two authors (HQ Gao and LN Li) independently evaluated the quality of the articles. Discrepancies were resolved by reaching a consensus with the senior author (JY Jing). The quality of the literature was evaluated using the Newcastle-Ottawa Scale (NOS) . The scaThe two authors (HQ Gao and LN Li) performed the risk assessment of the included studies. Any disputes or inconsistencies were discussed in the group to achieve a consistent result. The risk of bias was evaluated by the ROBINS-I (Risk Of Bias In Nonrandomized Studies—of Interventions) tool . Based oQ test and I2 test. When I2 was 0–40%, it means littler no heterogeneity, 30–60% means moderate heterogeneity, 50–90% indicates substantial heterogeneity, and 75-100% indicates considerable heterogeneity [The relationship between Helicobacter pylori and esophageal cancer was summarized by the binary method. OR and 95% CI were calculated for each study. To investigate the sources of heterogeneity, we carried out the following tests: heterogeneity tests, subgroup analysis, metaregression analysis, and sensitivity analysis. The heterogeneity of each study was statistically analyzed by ogeneity . If therogeneity . The ranogeneity , and subAccording to the inclusion and exclusion criteria, a total of 35 studies were included in our study , 26–56. According to the ROBINS-I tool, four studies had a serious risk of bias , 37, 46,P < 0.001). When analyzing population-based studies, the correlation coefficient between Helicobacter pylori infection and the risk of ESCC was 0.93 (95% CI: 0.68-1.28). Based on clinical studies, with a pooled OR of 0.66 (95% CI: 0.40-1.07). There were significant differences between the groups (P = 0.005). After grouping based on the detection method of Helicobacter pylori, the pooled OR of the detection group with various methods was 1.32 (95% CI: 0.87-2.01). There were significant differences between the groups (P = 0.003). According to the HKSJ method, Helicobacter pylori did not increase the risk of esophageal squamous cell carcinoma in North America . There was no statistically significant difference between the remaining comparative according to the DL methods and the HKSJ method.  and the I2 statistic had a higher heterogeneity (I2 = 78.5%). Hence, we performed a subgroup analysis to further explore the potential source of heterogeneity. The heterogeneity did not decrease significantly when grouped according to ethnicity, study object, and matching, but it decreased significantly when grouped according to the study area, study type, and Helicobacter pylori detection method. The results of the sensitivity analysis showed that after excluding one study, the pooling effect of the remaining studies was basically the same as the original total pooling effect. This confirmed that our results were stable.In the study of esophageal squamous cell carcinoma, the P values were 0.920, which indicated that there was no significant publication bias in the whole study.The results of the funnel plot showed no significant asymmetry , which sP < 0.001). After grouping based on the detection method of Helicobacter pylori, the OR of the detection group with various methods was 0.81 (95% CI: 0.32-2.07). There were significant differences between the groups (P < 0.001). Besides, the risk of Helicobacter pylori infection and esophageal adenocarcinoma was also negatively correlated after grouping based on ethnicity and study object; there were significant differences between the groups (P < 0.001). According to the HKSJ method, the pooled OR of the nonmatching group was 0.21 (95% CI: 0.13-0.36), while the OR of the matching group was 0.71 (95% CI: 0.60-0.84).  and the I2 statistic had a higher heterogeneity in the study results (I2 = 73.4%). The heterogeneity did not decrease significantly when grouped according to the study area and study type, but it decreased significantly when grouped according to ethnicity, study object, matching, and detection method of Helicobacter pylori. Besides, we did not find substantial changes in the corresponding pooled ORs by sensitivity analysis. This confirmed that our results were stable.In esophageal adenocarcinoma, the P value was 0.222, which indicated that there was no significant publication bias in the whole study.The results of the funnel plot showed slightly asymmetric signs , which sP = 0.135), ethnicity , study area , study type (P = 0.510), study object (P = 0.226), matching (P = 0.959), and detection method  were assessed independently. In esophageal adenocarcinoma, year (P = 0.022), ethnicity , study area , study type (P = 0.245), study object (P = 0.512), matching (P = 0.073), and detection method  were also assessed independently. The results of the univariate analysis were presented in P = 0.455, Europe: P = 0.368, East Asia: P = 0.395, Middle East: P = 0.05) and detection method  were assessed simultaneously. In esophageal adenocarcinoma, year (P = 0.49), ethnicity , and matching (P = 0.209) were also assessed simultaneously. The results of the multivariate analysis were presented in In order to further study the effects of these characteristics on estimating the relationship between esophageal cancer and Helicobacter pylori, we conducted the metaregression analysis. Year, ethnicity, study area, study type, study object, matching, and Helicobacter pylori detection method were entered as explanatory covariates. First of all, univariate metaregression analysis was performed. Regarding esophageal squamous cell carcinoma, in univariate metaregression analysis, year  and cyclooxygenase- (COX-) 2 in esophageal epithelial cells and plays a role in the inflammation associated with Barrett's esophagus and tumorigenesis in the esophagus [P < 0.001). The incidence of esophageal adenocarcinoma increases with age, and there is male predominance in esophageal adenocarcinoma. The incidence of esophageal adenocarcinoma in men and women is 6 : 1 in general, which is as high as 8 : 1 in some of the other populations, such as in the United States [As for esophageal adenocarcinoma, we found that Helicobacter pylori infection may reduce its risk, which is consistent with several previous meta-analyses , 15, 65.sophagus . Studiessophagus . When mad States , 67. Thed States  conducted States . Moreoved States .Heterogeneity might affect the interpretation of the results. This meta-analysis showed that there was considerable heterogeneity. Regarding esophageal squamous cell carcinoma, we performed a subgroup analysis of six possible factors. It was found that when grouped according to the study area, study type, and detection of Helicobacter pylori, heterogeneity decreased significantly. We then conducted metaregression and found that the study area may contribute more to the overall heterogeneity. As we have found, the risk of esophageal squamous cell carcinoma varies from area to area. In esophageal adenocarcinoma, subgroup analysis demonstrated that the heterogeneity was significantly lower after grouping by ethnicity, study object, matching, and detection of Helicobacter pylori. Metaregression showed that matching might contribute little to the overall heterogeneity; then, year and ethnicity may contribute more to the overall heterogeneity. However, multivariate regression analysis showed that year and ethnicity did not significantly differ. This may be related to the annual concentration of the articles we have included based on ethnicity.The advantages of our meta-analysis are as follows: Firstly, compared to the previous meta-analyses, our study included cohort studies, and the number of enrolled studies was more reliable. Subgroup analysis and metaregression were conducted to identify potential sources of heterogeneity; then, this study analyzed whether the differences between the groups were statistically significant. Secondly, all the studies in our meta-analysis have acceptable quality, the results of sensitivity analysis are stable, and the impact of publication bias is small. Thirdly, we analyzed the relationship between Helicobacter pylori infection and the risk of esophageal cancer in more details, whether different ethnicities and regions are the same and whether the differences have statistical significance. However, this study still has limitations. Firstly, this meta-analysis has not been registered online, which may cause bias in the analysis. Secondly, a special disadvantage of most original studies is the confounding factor; that is, many factors are not taken into account in research design or data analysis. Moreover, current epidemiological studies provide uncertain data on the negative or neutral correlation between Helicobacter pylori infection and esophageal adenocarcinoma. The analysis of observational studies supports the negative correlation that we have not analyzed the differences. Finally, the heterogeneity of the total combined effects of esophageal squamous cell carcinoma and esophageal adenocarcinoma is high in our study. Although these heterogeneities can be reduced after subgroup analysis, they might lead to a reduction in the number of studies and thus limit the reliability of data.In summary, no significant association was found between Helicobacter pylori infection and the risk of esophageal squamous cell carcinoma in the general population. However, lower risk was found in the Middle East when grouped by the study area. In the ethnic stratification analysis, there was no significant correlation between Helicobacter pylori infection and the risk of esophageal squamous cell carcinoma. Helicobacter pylori infection may reduce the risk of esophageal adenocarcinoma in the general population, but this may be one-sided; the statement of “protection effect” may be overestimated. Therefore, well-designed prospective cohort studies with a powered sample size are required, in which potential confounders should be taken into account to validate their relationship.


Recently, increasing reports showed that the risk of fracture may be correlated with type 2 diabetes mellitus (T2DM). However, their results still remained controversial. Thus we performed a meta-analysis including 11 studies to estimate the risk factor of limb fracture in type 2 diabetes mellitus.Databases including PubMed, Embase, Cochrane Library and Web of Science were searched to September, 2017. Risk Ratio (RR) with its 95% confidence intervals (CI) was used to evaluate the association between risk of limb fracture and type 2 diabetes mellitus. Two reviewers assessed the quality of all the included studies and extracted data for analysis independently.P = 0.0006).A total of 11 studies including 663,923 participants were included in this meta-analysis. Our analysis results showed that patients with type 2 diabetes mellitus had a significant association with risk of limb fracture , including leg or ankle fracture . Subgroup analysis showed individuals with type 2 diabetes had almost two-fold excessive risk of leg/ankle fracture in women and the pooled RR of leg/ankle fracture was 2.03 (95% CI 1.36–3.05; The results of the present meta-analysis showed that individuals with type 2 diabetes mellitus had higher risk of limb fractures, and this relationship is more pronounced in leg or ankle fracture. Fractures account for a large proportion in the global disease burden. At present, one of the most common risks of fracture should be osteoporosis, which lead a significant mortality, morbidity and socioeconomic burden in elderly patients . HoweverAt present, many studies have showed that diabetes mellitus patients usually experienced more opportunity of hip, pelvis, proximal humerus, distal forearm, ankle, knee, foot, wrist and vertebral fractures –25. Howe673 studies were retrieved after duplicated records removed from the preliminary screening of 1220 titles. By screening the titles and abstracts of the retrieved records, 17 studies of full texts were considered for further evaluation after 656 of records excluded. Six full articles were excluded at the final stage of filtrating work for reasons: two for review studies , 22, oneP = 0.02). The analysis was estimated using a random-effect model for the significant heterogeneity .There were eight studies providing data for estimating the association between risk of limb fractures and type 2 diabetes mellitus. As displayed in Figure P = 0.01). No significant results were observed in subgroups of humerus , forearm , wrist/hand/foot  and other limb fractures . Considering the significant intergroup heterogeneity, a random effect model was used.Subgroup analysis was conducted to find which site of fracture had significant association with type 2 diabetes mellitus. As shown in Figure p < 0.00001, I2 = 68%), a random effect model was used to combine the results.There were four studies providing data for estimating the association between risk of limb fractures and type 2 diabetes mellitus in women. As displayed in Figure P = 0.0006). No significant results were observed in subgroups of humerus , forearm , wrist/hand/foot .We also conducted subgroup analysis to find which site of fracture had significant association with type 2 diabetes mellitus in women. The results of subgroup analysis showed individuals with type 2 diabetes had almost two-fold excessive risk of leg/ankle fracture in women and the pooled RR of leg/ankle fracture was 2.03 , a fixed-effect model was used to combine the results. The results of subgroup analysis showed individuals with type 2 diabetes had an less risk of humerus fracture in women and the pooled RR of humerus fracture was 0.33 (95% CI 0.15–0.74). No significant results were observed in subgroups of leg/ankle fracture , forearm , wrist/hand/foot .There were three studies providing data for estimating the association between risk of limb fractures and type 2 diabetes mellitus in men. As displayed in Figure In order to examine the stability of the combined results, we performed sensitivity analysis. We conducted sensitivity analysis by omitting any single study then combining the rest of studies in the list. Sensitivity analysis showed that the combined RR of the association between risk of leg/ankle fracture and overall type 2 diabetes mellitus had high stability by omitting any single study. High stability was also found in the pooled result of risk of leg/ankle fracture in overall T2DM individuals. Sensitivity analysis indicated that there was none of any single study affecting the significance of combined RR of risk of leg/ankle fracture in T2DM individuals. However, for the association between risk of limb fractures and type 2 diabetes mellitus in men, sensitivity analysis showed that the study of Heath(2) et al. 1980 significantly affected the results of both overall group and subgroup, for that the pooled RR of overall group ranged from 0.77 (95% CI 0.63–0.94) to 0.86 (95% CI 0.68–1.08) and the pooled RR of subgroup of humerus ranged from 0.33 (95% CI 0.15–0.74) to 0.40 (95% CI 0.13–1.16) after omitting the study of Heath(2) et al. 1980. Funnel plots were conducted for assessing the publication bias of included literatures and we could roughly assess the publication bias by seeing whether their shapes were of any obvious asymmetry. The funnel plots showed no clear evidence of publication bias Figure In recent years, diabetes increasing the risk of fracture was increasingly given great interest. Up to the present, a great many studies have showed that diabetes mellitus patients usually experienced more opportunity of fracture in hip, pelvis, proximal humerus, distal forearm, ankle, knee, foot, wrist and vertebral. However, the mechanism between risk of fracture and diabetes has not been well studied to date. What’ more, it seems that the affection of diabetes is inequable in different site of fracture in different gender. For example, in study of Holmberg AH et al. , diabeteP = 0.02). Subgroups analysis further showed that individuals with type 2 diabetes had an excessive risk of leg/ankle fracture and the pooled RR of leg/ankle fracture was 1.80 . Similar conclusion was also found in women. However, opposite results were found in male risk of limb fracture. To find the cause of this phenomenon, sensitivity analysis was conducted. Sensitivity analysis found that one study, Heath(2) et al. [Thus, the present meta-analysis we performed for purpose of demonstrating the association between type 2 diabetes mellitus and risk of limb fractures. Eleven studies including 663,923 individuals were included in this meta-analysis and we found two case–control studies and nine cohort studies. These studies involved hip fracture, pelvis fracture, proximal humerus fracture, distal forearm fracture, ankle fracture, knee fracture, foot fracture, wrist and vertebral fracture. To well study the association between limb fracture and T2DM, we only extracted and took advantage of data and information about humerus, forearm, ankle, knee, foot, hand, leg and wrist fracture. In addition, in order to clear which site of limb fracture had significant association with type 2 diabetes mellitus, we performed subgroup analysis. Our analysis results strongly supported the conclusions that individuals with type 2 diabetes mellitus had a higher risk of limb fracture, with the pooled RR being 1.18  et al. , it was Several limitations for the present meta-analysis exist. The main limitation was due to the paucity of reported data. This meta-analysis included two case–control studies and nine cohort studies and both of them were retrospective observational researches. The majority of the studies we included were prospectively designed to measure the risk of fractures except the study of Schwartz AV, et al. 2001 . AdditioIn conclusion, the results of the present meta-analysis showed that individuals with type 2 diabetes mellitus had excessive risk of limb fractures, and this association was more significant in leg or ankle fracture, especially in female diabetes mellitus.The including criteria of this meta-analysis were as follows: (1) Both cohort and case-control studies were included; (2) Studies about the association between T2DM and fracture risk ; (3) RR with its 95% CI of association between T2DM and fracture risk were reported or can be obtained from studies.Excluding criteria were as follows: (1) Trials on animals; (2) Abstracts, letters, editorials, expert opinions, reviews, conference records, case reports; (3) Participants included in studies having other disease which could result in fracture or could increase risk of fracture; (4) Studies without sufficient data; (5) Duplicate articles were excluded.We searched PubMed, Embase, Cochrane Library and Web of Science to September, 2017. We also searched the citation lists of included studies. Our searching terms and procedures were as follows: “fracture” AND “diabetes” AND “risk factors”. We searched the databases with these terms in English, including references of some literatures we read. Two assessors independently screened the titles and abstracts of each study. Once relevant studies became certain, the full texts were obtained for further evaluation.Two reviewers assessed the quality of all the included studies using the 9-star Newcastle-Ottawa Scale (NOS) independently, and the total scores of each study were displayed in the characteristics table. The scores were judged according to the three aspects of NOS of evaluation: selection, comparability, and outcome between the case group and control group .Data for the analysis were extracted independently by two reviewers, and disagreement was resolved by their discussion. In addition, the extracted contents included study demographics, published years, country, trial design, cancer location, outcomes, using a standardized form. This study was performed strictly abiding by the standards of the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) .Data collected were input into Review Manager 5.2 software for analysis .In the present meta-analysis, the RR was used to evaluate the association between T2DM and risk of limb fracture. The associated 95% confidence intervals (CI) were also measured.Q statistical test, with P value and I2 statistic, ranging from 1 to 100%, to quantify the effect of heterogeneity [P ≤ 0.10 was deemed to represent significant heterogeneity [P ≥ 0.10), a fixed effects model . On thel method ) was useIn addition, we performed subgroup analysis according to site of fracture like proximal humerus, distal forearm, leg or ankle, wrist/hand/foot and other limb fractures. Besides, sensitivity analysis was performed to examine the stability of the combined results. Finally, publication bias was assessed by contour-enhanced funnel plots. If the shape of funnel plots revealed no obvious evidence of asymmetry, we considered that there was no obvious publication bias. All statistical analyses were performed using standard statistical procedures provided in Review Manager 5.2 .


Background: The available evidence on the effects of daylight saving time (DST) transitions on major cardiovascular diseases is limited and conflicting. We carried out the first meta-analysis aimed at evaluating the risk of acute myocardial infarction (AMI) following DST transitions. Methods: We searched cohort or case-control studies evaluating the incidence of AMI, among adults (≥18 y), during the weeks following spring and/or autumn DST shifts, versus control periods. The search was made in MedLine and Scopus, up to 31 December 2018, with no language restriction. A summary odds ratio of AMI was computed after: (1) spring, (2) autumn or (3) both transitions considered together. Meta-analyses were also stratified by gender and age. Data were combined using a generic inverse-variance approach. Results: Seven studies  were included in the analyses. A significantly higher risk of AMI  was observed during the two weeks following spring or autumn DST transitions. However, although AMI risk increased significantly after the spring shift , the incidence of AMI during the week after winter DST transition was comparable with control periods . No substantial differences were observed when the analyses were stratified by age or gender. Conclusion: The risk of AMI increases modestly but significantly after DST transitions, supporting the proposal of DST shifts discontinuation. Additional studies that fully adjust for potential confounders are required to confirm the present findings. Since decades, it has been shown that light-dark alternation influences the synchronization of circadian rhythms of most human systems . CellulaThe effects of time zone transitions (jet lag) are well known  and someMedLine and Scopus databases were initially searched to identify studies that evaluated the incidence of acute myocardial infarction (AMI) during the week after (a) spring transition into DST and/or (b) autumn switch back from DST to standard time, versus the incidence of AMI during control periods . The bibliographic search was performed by two independent investigators  up to 31 December 2018, using the following search terms: (daylight saving time (Title/Abstract) OR DST (Title/Abstract)) OR (circadian rhythm (Title/Abstract) OR circadian misalignment (Title/Abstract)) OR (sleep (Title/Abstract) OR sleep deprivation (Title/Abstract)) OR (chronobiology (Title/Abstract)) AND  OR AMI (Title/Abstract)). While maintaining a common overall architecture, several alternative strings were used. The reference list of the reviews and retrieved articles were also searched for additional pertinent papers. There was no language restriction.Studies were included if they: (a) had a cohort or case-control design; (b) included adult patients (≥18 years) with a diagnosis of AMI that was made in the first week after the shifts to and/or from DST and was documented through clinical chart, hospital discharge abstract databases or nationwide/local registries of pathologies; (c) provided enough data to compare AMI incidence before and after the time shifts.Individual study quality was assessed using an adapted version of the Newcastle Ottawa Quality Assessment Scale . We evalData were combined using a random-effect generic inverse variance approach , which eThe units of the meta-analysis were single comparisons of the rate of AMI in the first week following: (1) spring transition into DST; (2) autumn transition from DST; (3) both transitions considered together, versus the rate of AMI occurred during control weeks. Each of the above meta-analysis was also stratified by gender  and by age (<65 and ≥65 years). Thus, a total of 15 separate meta-analyses were performed.2 statistic [Between-study heterogeneity was quantified using the Itatistic . The pottatistic . Howevertatistic . We wereAll meta-analyses were performed using RevMan software, version 5.3 . Stats Direct, version 3.1, was used to compute Egger’s test .Among the 2633 retrieved papers, we identified seven studies  that met the selection criteria and were included in the analyses ,21,22,23The methodological characteristics of the included studies are summarized in p = 0.01).In the analyses considering both shifts to and from DST, each study contributed with two datasets (one for the spring and one for the autumn transitions), thus a total of 14 datasets  were included into the overall meta-analysis  17,18,1,21,22,23n = 10,382) and eight datasets [n = 33,587) were included in the meta-analyses assessing AMI risk for females and males, respectively  [n = 17,284) [In the analyses stratified by gender, three studies ,19,20 redatasets ,18,19,20ectively . The risectively : compare 15,525) ,19, as wp < 0.001). In the analyses stratified by gender or age, no other significant difference emerged, with the only exception of the meta-analysis restricted to the subjects aged ≥65 years: OR = 1.07; 95% CI: 1.00–1.14 , the funnel plot did not appear to be skewed and the corresponding Egger weighted regression tests did not identify asymmetry p = 0.27—.The main findings from this meta-analysis, based on seven studies including more than 100,000 participants, can be summarized as follows: (a) a significantly higher risk of AMI was observed in the weeks following spring and autumn DST transitions; (b) when spring and autumn DST shifts were considered separately, the risk of AMI increased significantly only in the week after the spring shift; (c) the increase of AMI risk, although significant, was relatively modest ; (d) in stratified analyses, which are likely affected by a lack of power, no substantial differences were observed by gender or age.As the main potential explanation for the observed findings, DST transition may cause a disruption of the circadian rhythm, which in turn induces changes in sleep quantity and quality , togetheFor most individuals, the alignment to the new time takes up to seven days ,34, althDespite the fact that about 1.6 billion people experience DST worldwide , with a Several limitations should be considered in the interpretation of the results. First, most of our meta-analyses showed an intermediate-to-high level of heterogeneity. However, a certain degree of heterogeneity across studies may not be surprising given the large variation in terms of setting and baseline patients characteristics, which is typical of meta-analyses of observational studies . Also, wThis meta-analysis showed a modest but significant increase in the risk of AMI following DST transitions, that was particularly noticeable after the spring DST shift. Overall, these findings support the proposal of a DST transition discontinuation, although additional evidence is certainly needed to confirm the present results, identify high-risk subjects and quantify the relationship between DST transition and other severe diseases.


The incidence of severe hand, foot, and mouth disease (HFMD) is not low, especially in mainland China in almost every year recently. In this study, we conducted a meta-analysis to generate large-scale evidence on the risk factors of severe HFMD to provide suggestions on prevention and controlling.  PubMed, Web of Science, Embase, Cochrane Library, China National Knowledge Infrastructure (CNKI), and Wanfang (Chinese) were searched to identify relevant articles. All analyses were performed using Stata 14.0.  We conducted a meta-analysis of 11 separate studies. Fever (odds ratio (OR) 7.396, 95% confidence interval (CI) 3.565–15.342), fever for more than 3 days , vomiting , limb trembling , dyspnea , contact with HFMD children , rashes on the hips , pathologic reflexes , Lethargy , convulsions , and EV71 infection  were significantly related to the risk of severe HFMD. We did not find an association between female sex , scatter-lived children , floating population , rash on the hands , rash on the foot , the level of the clinic visited first (below the country level) , breast feeding , and the risk of severe HFMD.  Fever, fever for more than 3 days, vomiting, limb trembling, dyspnea, contact with HFMD children, rashes on the hips, pathologic reflexes, lethargy, convulsions, and EV71 infection are risk factors for severe HFMD. Hand, foot, and mouth disease (HFMD) is a common childhood infection disease with characteristic features of fever, oral ulcers, and vesicular rashes on the hands, feet, and buttocks. It is caused by a group of enteroviruses, commonly coxsackie A16 and enterovirus-71 (EV-71). The mode of transmission of HFMD is mainly via the fecal-oral route, respiratory droplets, contact with blister fluid of an infected individual, or general close contact with infected individuals. Most HFMD cases were mild and limited to fever and vesicular exanthema on patients' palms, soles, and mouth along with discomfortness at certain levels. However, some severe cases with potentially fatal complications such as brain stem encephalitis (BE) and/or pulmonary edema (PE) show rapid progression that may lead to serious sequelae, even death. In recent years, more and more outbreaks of severe cases have been reported . PreviouA systematic search of the literature was done in the following electronic databases: PubMed, Web of Science, Embase, Cochrane Library, China National Knowledge Infrastructure (CNKI), and Wanfang (Chinese). The following keywords were used: severe hand foot mouth disease (HFMD), risk factors, and case control study. The quality of the included studies was assessed using the Newcastle-Ottawa Scale, and studies achieving six or more points were considered to be of high quality.Two investigators searched the electronic databases independently according to the following criteria for inclusion: (1) a case-control study including severe and mild disease patient groups; (2) published up to June 2017; (3) diagnosis of severe HFMD and mild HFMD consistent with the criteria defined by us. Abstracts, reviews, case reports, noncomparative studies, and low-quality studies were excluded. In cases of disagreement, a third investigator acted as an arbitrator, and the disagreements were resolved with the research team by discussion.The following items were extracted from the included studies: the first author's name, year of publication, source of publication, type of the study, risk factors, total sample size, number of severe and mild HFMD cases, and diagnostic criteria for severe and mild HFMD.P > 0.05, the publication bias exists; otherwise, the publication bias does not exist.The publication bias was evaluated using Egger's test . If P > HFMD cases were divided into two groups according to Guidelines on the Diagnosis and Treatment of HFMD . The milX2 and I2 statistics. I2 >50% and X2-statistic (P < 0.1) were considered to show significant heterogeneity, and the random effect model was adopted; otherwise, the fixed effect model was used. The OR and 95% CI were used as summary statistics for the comparison of the following risk factors: fever, fever for more than 3 days, vomiting, limb trembling, dyspnea, contact with HFMD children, rashes on the hips, pathologic reflexes, lethargy, convulsions, EV71 infection, female sex, scatter-lived children, floating population, rash on the hands, rash on the foot, the level of the clinic visited first (below the country level), and breast feeding.Stata 14.0 was used for the statistical analysis. The odds ratio (OR) and 95% confidence intervals (CI) were calculated using the fixed effect model or random effect model, and the choice for statistical model was determined by their heterogeneity which were assessed by the P values were 2-sided. A P value less than 0.05 was considered to be statistically significant.The pooled estimate of risk was obtained by the Mantel–Haenszel method in the fixed effect model and by the M-H heterogeneity method in the random effect model. All The first search strategy generated 109 studies. Only 11 articles  met the In 8 studies, fever was strongly related to the risk of severe HFMD . In 3 studies, fever for more than 3 days was significantly associated with severe HFMD . In 3 studies, rashes on the hips or buttocks were significantly associated with severe HFMD . However, we found no significant association between rash on the palm, rash on the soles, and severe HFMD in 3 studies, respectively  , limb trembling , dyspnea , pathologic reflexes , lethargy , and convulsions  were significantly associated with severe HFMD , scatter-lived children , and floating population  were all not related to the risk of severe HFMD –15.Seven studies analyzed the association between EV71 infection and severe HFMD, and the results suggest that EV71 infection significantly increased the probability of severe HFMD  .Seven studies suggest that contacting with HFMD children significantly increased the risk of severe HFMD . We found no association between the level of the clinic visited first (below the country level) , breast feeding , and severe HFMD –19.Egger's test analysis of total complications was performed. The results are shown in A recent meta-analysis involving 19 separate studies  found thPrevious studies on gender have different conclusions: Pan et al.  thought Both male and female have the same opportunities to develop severe HFMD. Previous studies have also drawn different conclusions regarding scatter-lived children and floating populations: Zhang et al.  thought Oral ulcers and vesicular rashes on the hands, feet, and hip/buttocks are common signs of HFMD; however, rashes on different parts can lead to different levels of severity of HFMD. Previous studies regarding rashes on the hands and rashes on the hips have different conclusions: Yang et al.  thought Most of previous studies included in our meta-analysis found that fever, vomiting, lethargy, convulsions, and contact with HFMD children are risk factors of severe HFMD, and our study confirmed it. Our study also confirmed that severe HFMD is associated with fever for more than 3 days, limb trembling/shaking, dyspnea, pathologic reflexes, and EV71 infection again. Any of these factors increase the possibility of developing severe HFMD. Therefore, early recognition and meticulous management of patients with these risk factors are required , 23.EV71 invades the central nervous system causing severe disease ranging from meningitis to fatal encephalitis . In our In recent years, coxsackievirus A6 (CV-A6) has caused widespread concern around the world and has gradually emerged as a major pathogen of the hand-foot-mouth disease (HFMD) . The infLiu et al.  reportedIn conclusion, we found that eleven factors are associated with the severity of HFMD. Previous conclusion regarding the association between fever (body temperature ≥37.5°C), fever more than 3 days, lethargy, vomiting, and EV71 infection and severe HFMD was consistent with our findings. Also, we found rashes on the hips or buttocks, limb trembling/shaking, dyspnea/breathlessness, pathologic reflexes, convulsions/twitch, and contact with HFMD children significantly increased the risk of severe HFMD. But, we found no significant association between rashes on the palm, rashes on the soles, female gender, scatter-lived children, floating population/migrant, the level of the clinic visited first (below the country level), breast feeding, and severe HFMD. Further studies are needed to confirm our findings.


Pancreatic cancer is one of the most malignant and aggressive cancers, with poor survival rates and diagnosis being based on nonspecific tumour biomarkers. Due to this, the publication, “Systematic Review and Meta-Analysis of Diagnostic Accuracy of miRNAs in Patients with Pancreatic Cancer”, by Sun et al. has excellent potential for highlighting potential miRNAs as high-accuracy diagnostic markers in pancreatic cancer . HoweverThe authors have stated that there is a need to measure the diagnostic value of miRNA in pancreatic cancer. However, previous studies in the field already exist –4. DespiAnother issue is that sensitivity and specificity did not differ across all the included studies. This is applicable only in a scenario where all the studies included in the pooled meta-analysis have a singular diagnostic cut-off point . HoweverI-square parameters may not be sufficiently informative as they ignore the threshold effect. As this study follows a random effects model (due to the presence of between-study heterogeneity), the tau-squared statistical parameter, being the estimated variation of heterogeneity between the effects for test accuracy observed in different studies, might be suitable for inclusion in this meta-analysis study. Furthermore, the authors have also stated that Deeks et al.'s test performed showed a statistically nonsignificant value. It is important to consider that Deeks et al.'s test is used to identify the effective same sample size, and it is insufficient in clarifying the presence of patient-level or study-level bias that exists [In the statistical analysis as well as in the subsequent interpretation of results, the chi-square and t exists .This study would be a valuable resource for future studies in this field if these issues were to be addressed.


C3 gene account for a significant risk of advanced AMD.The purpose of this study is to discuss whether genetic variants  in the We performed a meta-analysis using electronic databases to search relevant articles. A total of 40 case-control studies from 38 available articles  were included in our study.P < 0.001; OR = 1.45, 95% CI = 1.37–1.54, P < 0.001). Moreover, in the subgroup analysis based on ethnicity, rs2230199 and rs1047286 polymorphisms were more likely to be a predictor of response for Caucasian region . Besides, pooled results suggested that the G allele of rs2230199 could confer susceptibility to advanced AMD in Middle East .In our meta-analysis, the pooled results showed that the carriage of G allele for rs2230199 and the T allele for rs1047286 had a tendency to the risk of advanced AMD  contains supplementary material, which is available to authorized users. Age-related macular degeneration (AMD) is a complex and progressive retinal disorder influenced by family history, aging, race, smoking and diet, which caused irreversible visual impairment in a growing number of elderly persons , 2. The CFH) gene by candidate region and whole genome association analyses [Advanced AMD has been implicated with important risk factors listed above, it is a multifactorial disease which influenced by a combination of environmental and genetic susceptibility , 3, 7, 8analyses , 13.C3 gene have been investigated with advanced AMD in different ethnic groups, the pooled results are incompatible and ambiguous. According to the International HapMap Project database, the human C3 gene is located on chromosome 19 and exhibits nine common genetic SNPs . Used for screening the electronic database and manual searching, the most widely condidate polymorphisms of the C3 gene which at least has been surveyed in three pertinent studies are rs2230199, rs1047286, rs2230205 and rs2250656. In order to better understand the genetic risk of C3 gene in the relationship with exudative AMD, we performed a meta-analysis to illuminate this association and determine whether the genetic variants of C3 gene conferred susceptibility to advanced AMD.Apart from CFH, the central element of the complement cascade, complement component C3a has been interconnected with the vascular endothelial growth factor expression, geographic atrophy, retinal pigment epithelium deterioration, and progression to choroidal neovascularization , 14, 15.C3” or “C3” or “complement component 3”) and (“variant” or “mutation” or “genetic” or “SNP” or “polymorphism” or “genetic polymorphism” or “genetic variant” or “single nucleotide polymorphism”). Each database was thoroughly scanned and was up to date as of September 1 2018. Our meta-analysis was mainly focused on case-control studies, without any language limitation imposed in the literature searching.A systematic search of electronic database such as PubMed, Embase, CNKI, Cochrane library and Web of Science was conducted with the following keywords: (“AMD” or “maculopathy” or “macular degeneration” or “age-related maculopathy” or “age-related macular degeneration”) and  detailed genetyping data for each site could be acquired to estimate the odds ratio (OR) and 95% confidence interval (CI) based on genetic model contrast; (3) individual for all selected samples met the modified version of the age-related eye disease study (AREDS) grading system as described elsewhere. Major exclusion criteria were limited to several items as follow: (1) overlapping subjects in several articles for the same research group; (2) only focused on families’ individuals rather than sporadic advanced AMD patients; (3) abstract from conferences, letters, review articles and case reports. When several articles included some of the same samples, the one with largest individuals and thorough genotype information would be winnowed for our meta-analysis.Retrieved articles were considered eligible for our meta-analysis when they met the following inclusion criteria: (1) investigating the disease risk of Data from the retrieved studies were extracted independently by two reviewers (J.Z. and S.L.). The following items obtained from each eligible articles included: the first author, the year of publication, country and ethnicity of subjects, information on study design, sample size, genotyping methods and distribution in case and control groups. Two authors carefully inspected the raw statistics and reached a consensus in all aspects. If any disagreement still existed, the third author (S.H.) would be invited to chew over current controversy and resolve the dispute.Quality assessment of the screened studies was also independently conducted by two reviewers (J.Z. and S.L.) in the basis of the HuGENet Handbook . A totalC3 polymorphic site were counted between cases and healthy controls. The genetic strength association including pooled ORs and 95% CIs was assessed using different genetic models, including allele model (A vs. a), homozygote model (AA vs. aa), heterozygote model (Aa vs. aa), dominant (AA+Aa vs. aa), recessive (AA vs. Aa+aa). The heterogeneity assumption between studies was estimated and evaluated by Cochran’s Q statistic as well as the I2 statistic. The result that our P value of Q statistic was less than 0.05 or the I2 value was greater than 50% suggested apparent heterogeneity, thus a random-effect model was utilized in our model analysis. Otherwise, the fixed-effect model was performed [P values was executed by STATA 12.0 software . A significant difference was estimated under the level of 0.05. The final results needed to be tested and verified by two authors (J.Z. and S.L.) respectively.Allele and genotype frequency of each erformed . Sensitierformed  and Eggeerformed  were furC3 genetic polymorphism. The other 46 full-text articles were left in our meta-analysis. Seven of them did not have detailed genotype data after cautiously reading the included literatures.Besides, two papers were investigated by the same author and the same batch of patients from Iran [C3 gene with advanced AMD from 38 available publications were generally contained in our current meta-analysis [The flow diagram for literature searching is summarized in Fig. rom Iran , 21. We analysis , 22–57. Overall results in Table C3 polymorphisms including rs2230199, rs1047286, rs2230205, rs2250656 were used in our meta-analysis and the combined results are presented in Table Several genetic models for P < 0.001; homozygote model: OR = 2.33, 95% CI = 1.98–2.74, P < 0.001; heterozygote model: OR = 1.53, 95% CI = 1.41–1.64, P < 0.001; dominant model: OR = 1.62, 95% CI = 1.51–1.74, P < 0.001; recessive model: OR = 1.99, 95% CI = 1.70–2.34, P < 0.001). Moreover, the subgroup analysis straitified by ethnicity indicated that rs2230199 conferred obvious susceptibility to advanced AMD in the group of Caucasian in allelic  , heterozygote , dominant , recessive  models (Table P < 0.001). However, this relationship was not significant in East Asian group for any genetic models . In subgroup analysis stratified by ethnicity, our meta-analysis indicated significant correlation of rs1047286 with advanced AMD in the five genetic models . Subgroup analysis of Caucasian and East Asian group showed that there was a lack of relationship in any of the genetic models . But a weakly protective risk between this SNP and advanced AMD was observed in heterozygote model and dominant model . In the stratified analysis by ethnicity, there was no association in any of the genetic models.   . Furthermore, as emerged in Table P values for both tests are more than 0.05.Publication bias is a potential problem, thus Begg’s funnel plots and Egger’s regression tests were applied to investigate the publication bias for C3 has been a plausible candidate gene since its cleavage product C3a was confirmed in drusen. In our current meta-analysis, 20,673 patients and 20,025 controls from 38 articles were combined to detect the association of C3 genetic polymorphisms with advanced AMD. We came to the conclusion that two nonsynonymous SNPs rs2230199 and rs1047286 were demonstrated an increased pathogenic effect on advanced AMD . Moreover, our meta-analysis discovered that SNP rs2250656 decreased the risk of advanced AMD susceptibility, which a protective association was acquired in heterozygote model and dominant model. Obviously, the results of SNP rs2250656 with advanced AMD needed to be validated with larger samples and studies in different ethnicity.AMD is a multifactorial disease, in which complement system mediated inflammation plays a pivotal role. Several pathways including the alternative complement component have been described to be implicated in the development of AMD . As the Being consistent with previous studies, the G allele of rs2230199 conferred susceptibility to advanced AMD in Caucasian group. In our meta-analysis, we first confirmed that the G allele of rs2230199 could be linked with AMD in Middle East but not East Asian region, though rather larger population needed to be validated in the future. Besides, our meta-analysis found a novel association between the T allele of rs1047286 and advanced AMD in Caucasian but not East Asian group.C3 gene have been identified as genetic risk factors for advanced AMD in Caucasian populations. However, the allele frequencies of rs2230199 vary widely among different ethnicities. Frequencies of the risk G allele at rs2230199 were 25% to 31% in AMD cases and 19% to 21% in controls in Caucasians [The common polymorphisms rs2230199 and rs1047286 in the ucasians . Besidesucasians , 57. Whiucasians . For rs1ucasians  also fouC3 is located on the short arm of chromosome 19 and consists of 41 exons, which forms 13 functional domains. C3 is the most abundant complement component and significant C3 messenger RNA is detected in the neural retina, choroid, RPE, and cultured RPE cells [C3 undergoes important structural rearrangements which causes conformational changes exposing binding sites for complement components and drusen including C3 and its activation products was confirmed in the finding that local inflammation and activation of the complement cascade can contribute to the pathogenesis of AMD. Notably, animal studies conducted by Bora et al. [C3 deficiency in C3−/− mice prevented the formation of choroidal neovascularization in advanced AMD (wet AMD), indicating that C3 is a pivotal element of this activation process.The gene for PE cells . CleavagPE cells  argued ta et al. , 61 haveC3F and C3S), the two alleles showed a differential capacity to bind monocyte- complement receptor. Helgason et al. [C3F) was associated with the reduction of C3 gene binding to CFH, which leads to an increase in complement activation. Additionally, rs2230199 variant may alter the net charge of the molecule and influence the position of the thioester-containing domain. Except for advanced AMD, the risk variant of rs2230199 has been previously considered as associated with other immune-mediated conditions, such as IgA nephropathy, systemic vasculitis.In our meta-analysis, four SNPs including rs2230199, rs1047286, rs223205 and rs2250656 were analyzed in the pooled data. Among them, rs2230199 and rs1047286 are located in the first ring of macroglobulin domains, which conduct a prominent function for correct orientation of the thioester-containing domain. The amino acid changes induced by the genetic mutations may alter the configuration of the macroglobulin ring . With evn et al.  noted th2 = 0.80), which haplotype analyses suggested that the effect of the C3 alleles was independent from the established genetic and environmental risk factors. Furthermore, our pooled analysis of neighboring SNPs of rs2230199 indicated that the allele frequency of the variant rs2230205 and rs2250656 was not significantly different between the advanced AMD cases and controls. Pei et al. [In the current meta-analysis, rs1047286 variant showed significant association with advanced AMD in Caucasian populations. Despriet et al.  argued ti et al.  confirmeIn a previous meta-analysis where a total of 15 independent studies with 5593 cases and 5181 controls were included, Zhang et al.  indicateC3 genetic polymorphisms and advanced AMD, some limitations for our present meta-analysis need to be declared. First, heterogeneity among the ethnic groups was discovered when investigating the association of C3 genetic variants with advanced AMD. However, based on the results of the sensitivity analysis, it is clear that the overall effect was not affected by heterogeneity. Additionally, there was no obvious publication bias detected in the contrast of C3 gene with advanced AMD. Second, the number of patients and controls was relatively small in each included study; therefore, a great number of samples from different ethnic regions are required for further analysis. Third, the effects of common confounding factors, including sex, age, body mass index, smoking, and diet were not evaluated in the present study because of insufficient data. Fourth limitation is that only three ethnic backgrounds with relatively few studies were taken into consideration, thus further efforts to reduce the incidence of ethnic bias will be needed once raw data become available. Finally, the electronic databases from which we selected eligible studies were listed in English and Chinese; therefore, a language bias may be existed in our meta-analysis.Considerable efforts have been paid to discuss the potential relationship between The present meta-analysis provided a series of evidence-based pooled data for a significant association between rs2230199, rs1047286 and susceptibility to advanced AMD, especially in Caucasians. Additional well-designed work with a larger number of studies in which incorporate different ethnicities together with gene-gene and gene-environment is recommended to better confirm the functional role of the two nonsynonymous polymorphisms.Additional file 1:The full details of databases searching terms. (DOC 38 kb)Additional file 2:Figures S1-S3. The sensitivity analysis of C3 genetic polymorphisms . (DOC 106 kb)


The association between interleukin-10- (IL-10-) 592  C>A polymorphisms and susceptibility to chronic or aggressive periodontitis (CP or AgP) is conflicting. This meta-analysis is aimed at quantitatively estimating the association.  PubMed, Embase, Web of Science, and WANFAN were searched for studies performed prior to January 31, 2018, to collect data for our research. Meta-analysis was performed using RevMan 5.3 or STATA 14.0.  In total, 18 studies that met our criteria were included. Overall or HWE subgroup analysis of individuals with this polymorphism revealed that in terms of CP susceptibility, there was a significant difference between case groups and control groups in the A allele versus C allele model , in the AA versus CC+CA model , and in the CC versus CA+AA model ; subgroup analysis based on a nonsmoking population also displayed significance in the A allele versus C allele model  and CC versus CA+AA model . For this polymorphisms and AgP susceptibility, our analyses revealed a significant association in both the A allele versus C allele model  and the AA versus CC+CA model ; subgroup analysis based on Caucasian or nonsmoking populations showed significant differences in the AA versus CC+CA model .  IL-10-592  A allele and the associated AA genotype may be risk factors for the onset of CP or AgP—particularly for the AA genotype and the increased risk of AgP in Caucasian or nonsmoking populations. Conversely, the CC genotype may act as a protective factor against the onset of CP. The [ATA] haplotype has been associated with decreased synthesis of IL-10 and is frequently associated with periodontitis . Owing tEven considering the limitations of this study, the present meta-analysis supported the hypothesis that IL-10-592  C>A polymorphisms may be associated with CP and AgP susceptibility. We not only identified that the IL-10-592  A alleles and AA genotypes may be a risk factor for the development of CP and AgP but also found out that the IL-10-592  CC genotype may play a protective role in preventing CP. It is noteworthy that the AA genotype was found to be more closely tied to the risk of AgP in Caucasian and nonsmoker population. Thus, IL-10-592  A alleles or AA genotypes may be a putative biomarker for diagnosing CP and AgP. Large-scale studies to further validate our findings should be performed in the future.


Aim We aimed to determine if nut consumption decreases mortality and/or the risk of cardiometabolic diseases based on updated meta-analyses of epidemiological and intervention studies. Methods. An updated electronic search was conducted in PubMed/MEDLINE, Cumulative Index to Nursing and Allied Health Literature (CINAHL), and the Cochrane Library databases for original meta-analyses to investigate the effects of nut consumption on cardiometabolic disease in humans. Results. Seven new meta-analyses were included in this updated review. Findings similar to our previous review were observed, showing that nut consumption significantly decreased cardiovascular disease (CVD) mortality , coronary heart disease (CHD) mortality , stroke mortality , CVD incidence , CHD [or coronary artery disease (CAD)] incidence , and stroke incidence  comparing high with low categories of nut consumption. Fasting glucose levels , total cholesterol , and low-density lipoprotein cholesterol  were significantly decreased with nut consumption compared with control diets. Body weight and blood pressure were not significantly affected by nut consumption. Conclusion. Nut consumption appears to exert a protective effect on cardiometabolic disease, possibly through improved concentrations of fasting glucose, total cholesterol, and LDL-C. Nuts comprise 43–67% fat and 8–22% protein by weight. Nuts have unique nutritional profiles. Nuts are abundant in unsaturated fatty acids (UFAs) (containing both monounsaturated fatty acids (MUFAs) and polyunsaturated fatty acids (PUFAs)) and only 4–5% of saturated fatty acids (SFAs). Nuts are high in vitamins, minerals, fiber, and bioactive compounds, including carotenoids, phytosterols, and polyphenols ,2.As we suggested, for a potential mechanism of action of nuts , MUFAs an = 6), cardiovascular disease (CVD) mortality , coronary heart disease (CHD) mortality , and stroke mortality . Moreover, nut consumption was associated with lowered risks of CVD , CHD , stroke , and hypertension  from meta-analyses of prospective studies. No effect of nut consumption on T2DM risk was observed in prospective studies, while significantly lowered fasting glucose levels (−0.08 to −0.15 mmol/L) were observed in meta-analyses of randomized control trials (RCTs). In meta-analyses of RCTs, decreases in total cholesterol  and low-density lipoprotein cholesterol  and improvements in endothelial function (0.79% to 1.03% increase in flow-mediated dilation) were observed. No effects on body weight, inflammatory markers, and blood pressure were seen with nut consumption.As previously reported , nut conThis review aims to update our previous review of meta-analyses in order to determine the effect of nut consumption on cardiometabolic disease.The literature search was conducted in PubMed/MEDLINE, Cumulative Index to Nursing and Allied Health Literature (CINAHL), and the Cochrane Library databases, restricted to full articles investigating meta-analyses of the effects of nut consumption on cardiometabolic diseases in humans. Only research articles written in English up to 11 November 2019 were included. The search terms included meta-analysis combined with nut(s) or tree nut(s) or almond(s) or Brazil nut(s) or cashew nut(s) or hazelnut(s) or macadamia(s) or peanut(s) or pistachio(s) or walnut(s) or mortality or incidence or CVD or coronary heart disease (CHD) or stroke or T2DM or hypertension or metabolic syndrome or obesity or blood pressure or glycemic control or glucose or lipids or inflammatory markers or endothelial function and flow-mediated dilation. Reference lists of chosen articles were also screened for related publications. A previous review  examined2 = 3%; pheterogeneity = 0.42) in a meta-analysis of 14 prospective studies  in a meta-analysis of 12 prospective studies  cications  includedications ,28.2 = 0%; pheterogeneity = 0.45) from a meta-analysis of 11 prospective studies  when 3 prospective studies . This publication [A previous review  reportedl., 2017 ) from thlication  was only2 = 74%; pheterogeneity < 0.01) in a meta-analysis of 7 prospective studies  for the highest versus lowest nut consumption. The RR per 28 g/d of nut consumption was 0.78 .In a meta-analysis of 3 prospective studies ,47,48 cop = 0.03) from a meta-analysis of 14 RCTs (12 publications [p = 1) was observed from a meta-analysis of 10 RCTs (9 publications [p = 0.72) was observed from a meta-analysis of 15 RCTs  from a meta-analysis of 23 RCTs . However, almond consumption did not affect body mass index , compared with controls from a meta-analysis of RCTs [Lee-Bravatti et al., 2019  conducte of RCTs ,78,79,86In a previous review , nut conIn this present review, we have included very recent meta-analyses conducted by Tindall et al., 2019 , and Sch2 = 53.4%) from a meta-analysis of 39 RCTs [2 = 51.0%). Tindall et al., 2019 [2 = 51.7%) and fasting insulin  after nut consumption from meta-analyses of 19 RCTs [Tindall et al., 2019 [ 39 RCTs ,116,117. 19 RCTs ,114,117 2 = 0%, pheterogeneity = 0.94) at >42.5 g almond consumption  compared with controls.Lee-Bravatti et al., 2019  conducteSchwingshackl et al., 2018  conducten = 450) by Viguiliouk et al., 2014 [A meta-analysis by Mazidi et al., 2016  that incl., 2014  showed aOverall, nut consumption appears to decrease fasting glucose levels by 0.08 to 0.15 mmol/L compared with control diets.In a previous review , nut conIn this present review, we included a meta-analysis conducted by Schwingshackl et al., 2018  and Lee-In a network meta-analysis of 66 RCTs ), Schwingshackl et al., 2018  reported2 = 67%; p < 0.01) and LDL-C (summary net change = −5.38 mg/dL (0.139 mmol/L); 95% CI −9.91, −1.75 mg/dL; I2 = 61%; p < 0.001) with almond consumption in a meta-analysis of 13 RCTs from 14 publications [2 = 71%; p < 0.01) following almond consumption compared with controls in a meta-analysis of 12 RCTs from 12 publications [Lee-Bravatti et al., 2019  found siications ,139,140 ications ,102,138 Overall, nut consumption lowers total cholesterol  and low-density lipoprotein cholesterol  compared with control diets in 10 meta-analyses ,136,137.2 = 49%, pheterogeneity = 0.08) between almond consumption and control for diastolic blood pressure (DBP) in the main meta-analysis of RCTs [2 = 35%, pheterogeneity = 0.21) and >6 weeks . Lee-Bravatti et al., 2019 [2 = 51%, pheterogeneity = 0.07) in summary estimates with almond RCTs [A previous review  showed n of RCTs ,80,81,86l., 2019  showed nond RCTs  comparedp < 0.001; I2 = 79.5%; pheterogeneity < 0.001). A subgroup analysis showed this inverse association was only present in tree nuts  but not in peanuts .Zhang et al., 2019  conducteThis present review is an updated review of meta-analyses that adds seven more recent meta-analyses aiming to clarify the effect of nut consumption on cardiometabolic disease. We found similar outcomes to our previous review when we combined new and previous meta-analyses investigating CVD mortality, CHD mortality, stroke mortality, CVD incidence, CHD (or CAD) incidence, and stroke incidence. In meta-analyses of interventions, nut consumption significantly reduced fasting glucose levels, TC, and LDL-C compared with controls. However, body weight and blood pressure did not differ after nut consumption compared with controls. Schwingshackl et al., 2018  showed tThe results from this present review suggest there needs to be further large clinical trials testing nuts as therapeutic agents for primary and secondary prevention of cardiovascular disease.This updated review of meta-analyses found that nut consumption has beneficial effects on cardiometabolic disease with reduced CVD mortality, CHD mortality, stroke mortality, CVD incidence, CHD incidence and stroke incidence comparing high with low categories of nut consumption. It may be attributable to decreases in fasting glucose, total cholesterol and LDL-C.


HSPG2) and chondroitin sulfate proteoglycan 2 (CSPG2) are associated with increased risk of intracranial aneurysms (IAs) susceptibility.We aimed to investigate whether the polymorphisms of gene heparan sulfate proteoglycan 2  with 95% confidence intervals (CIs) were presented to assess the associations between CSPG2  or HSPG2 , and a total of 8651 participations (3674 with IAs and 4977 for control). For the rs251124 polymorphism of CSPG2, the quantitative synthesis from 5 studies showed significant difference in the gene allele comparison of T vs. C . Similar results were found for rs3767137 of HSPG2 . However, for the rs173686 polymorphism of CSPG2 and rs7556412 polymorphism of HSPG2, no significant difference was found We enrolled 7 papers, 4 matched single nucleotide polymorphisms (SNPs) of CSPG2 and rs3767137 of HSPG2 had statistically significant associations with IAs susceptibility. The minor allele T of rs251124 demonstrated a harmful effect but the minor allele A of rs3767137 demonstrated a protective role with regard to the risk of IAs. However, no such associations were found in the SNPs rs173686 of CSPG2 and rs7556412 of HSPG2.The SNPs rs251124 of  Intracranial aneurysms (IAs) affect approximately 2% of the general population and account for 85% of subarachnoid hemorrhage (SAH) , 2. IA rVarious theories have been proposed for the development of IA, and IA is always considered as a multifactorial disorder due to the interaction of both environmental and genetic factors. Environmental factors, such as alcohol consumption, smoking, and hypertension seem to be common IA risk factors , 7. HoweHSPG2) and chondroitin sulfate proteoglycan 2 (CSPG2) genes have been identified as susceptibility genes for IA . The NOS employed a star rating system to assess quality from 3 broad perspectives of the study: 1) selection of the study groups, 2) comparability of the groups, and 3) identification of the exposure (for case-control studies) or outcome of interest (for cohort studies). Scores ranged from 0 to 9 stars, and studies with no less than 7 stars were considered to be of high quality.P-values (less than 0.05 was considered significant). I2 statistics, with the value ranged from 0% to 100% (complete consistency to complete inconsistency), was used to determine the statistical heterogeneity among studies. If the I2 -value was more than 50%, the random-effects model was chosen to calculate the pooled OR; otherwise the fixed-effects model was used. All of the results were presented as forest plots.The statistical analysis was performed with the software named ‘‘Comprehensive Meta-Analysis ’’. The association strength between the gene polymorphisms and IAs susceptibility was calculated by the Z test, presented with the OR (respective 95% CIs). And the significance of the pooled OR was determined with The initial literature search retrieved 87 relevant articles (duplicates were discarded), and finally, 7 studies were included in the present study after carefully review. Flowcharts describing the study selection was shown in Overall, 8651 participations (3674 with IAs and 4977 for control) were enrolled in the studs. All the patients were radiologically confirmed intracranial aneurismal cases and the controls were age and gender matched. CSPG2 and HSPG2 gene variants between patients and controls were extracted and analyzed from the included studies. Only the meta-analysis of the allele frequencies was able to conduct between the studies, the meta-analysis of the genotype frequencies was not available because of the limited data. The meta-analysis for the relationship between gene CSPG2  or HSPG2  polymorphisms and IAs susceptibility were shown in The genotype and allele frequencies of the CSPG2, the quantitative synthesis from 5 studies showed the gene allele significant difference in the gene allele comparison of T vs. C  .For the rs251124 polymorphism of  P=0.01) . HoweverHSPG2, quantitative synthesis showed significant differences in the comparisons of allele frequencies .Similarly, for SNPs rs3767137 of Egger's test was performed to access the publication bias of studies included in this meta-analysis. The Egger's test demonstrated no evidence of publication bias for all the four SNPs.CSPG2 and rs3767137 of HSPG2 had statistically significant associations with IAs susceptibility. The minor allele T of rs251124 demonstrated a harmful effect but the minor allele A of rs3767137 demonstrated a protective role with regard to the risk of IAs. However, no such associations were found in the SNPs rs173686 of CSPG2 and rs7556412 of HSPG2. Because of the increasing number of cases and controls in this analysis, the estimates of association obtained are more precise.The study demonstrated the SNPs rs251124 of CSPG2 and HSPG2 encode two important proteins in maintains of the ECM, thus the genotype variations of CSPG2 and HSPG2 and the resulting functional changes may be an important part of IAs pathologies. Until now, contrary results were reported with regard to the relationship between CSPG2 and HSPG2 polymorphisms and IAs susceptibility. In the Dutch population, SNPs in strong linkage disequilibrium and haplotypes constituting these SNPs in the CSPG2 gene were associated with IAs  have been completely observed by the authors.


Recently, we received the letter to the editor  in respoRegarding the differences between this study and previous publications –5, we wiThe diagnostic cut-off point plays an important role in disease diagnosis. There is not a singular diagnostic cut-off point for different miRNA profiling and source of miRNA. We originally prepared our subgroup analysis and sensitivity analysis based on different miRNA profiling and source of miRNA, but half of the included studies did not report the testing threshold. Therefore, such subgroup analysis and sensitivity analysis cannot be performed, which may have an impact on the results. It was recommended that diagnostic test articles should report thresholds to provide raw information for future meta-analysis.P = 0.127), suggesting that there is no threshold effect. Then an I-square parameter was adopted to estimate the heterogeneity between the studies. However, we agree with Jayaraj et al.'s suggestion that a Tau-squared statistical parameter might be suitable for being the estimated variation of heterogeneity between the effects for test accuracy observed in different studies. For the meta-analysis of diagnostic test accuracy (DTA), the hierarchical summary receiver operating characteristic (HSROC) allows both the existence of threshold effects and the heterogeneity between studies. We subsequently conducted HSROC and found that the pooled SEN was 0.82  and the pooled SPE was 0.79    and basiThe Cochrane Handbook for Systematic Reviews explicitly mentions not to use methods like the Begg or Egger tests for publication bias of DTA and argues that it is best to use the test proposed by Deeks . Applyin


Many studies have reported that the IL-1β + 3954C/T polymorphism (rs1143634) is related to myocardial infarction (MI). To classify the association between IL-1β + 3954C/T and MI susceptibility, we performed a meta-analysis.We retrieved relevant literature from electronic databases . Pooled odds ratios (ORs) and 95% confidence intervals (95% CIs) were calculated with a fixed effect model or a random effect model. Sensitivity analysis and publication bias results are also presented.I2 = 0%, PH = .448) and in a dominant model . Ethnic subgroup analysis showed similar results in Caucasian populations: an allelic comparison , homozygote model , and dominant model . In addition, similar effects remained in subgroups analyses of high-quality studies and PCR-RFLP (restriction fragment length polymorphism) data.Nine eligible studies (2299 controls and 2203 cases) were included. The pooled results showed a significant relationship between MI and IL-1β + 3954C/T in an allelic comparison (T vs C: OR = 1.13, 95% CI 1.02–1.25, Our meta-analysis proved that IL-1β + 3954C/T is associated with MI susceptibility, especially among Caucasian populations. Most MIs are related to coronary artery disease (CAD) and coronary atherosclerosis rupture is the most frequent cause of MI.,3 It is well known that inflammation plays a primary role in atherosclerosis and plaque rupture. Inflammation seems to influence each stage of atherosclerotic development, such as oxidative injury, cell proliferation, and plaque evolution and instability.,8Myocardial infarction (MI), a highly prevalent cardiac emergency caused by a vital disequilibrium between oxygen supply and demand in myocardial cells, is a primary cause of morbidity and mortality worldwide. IL-1β plays a key role in coronary atherosclerotic heart disease and atherosclerotic inflammation. Some studies have reported that the expression of IL-1β was elevated in the myocardium early after injury.,13 A single nucleotide polymorphism (SNP) was identified in exon 5 at position +3954C/T of the IL-1β gene. The T allele of IL-1β + 3954C/T is less common than the С allele and is related to an elevated serum IL-1β level. A polymorphism leading to IL-1β overproduction may increase the risk of autoimmune diseases such as atherosclerosis. Although many studies have presented a connection between IL-1β + 3954C/T and MI risk,–23 the sample sizes of these studies were limited, and the results are controversial. Thus, we performed a meta-analysis to clarify the association between IL-1β + 3954C/T and MI susceptibility.The pro-inflammatory cytokine interleukin-1 beta (IL-1β) is involved in the initiation of multiple biological cascades that are important parts of the inflammatory reaction.22.1A systematic search was performed in PubMed, Cochrane, Embase (Excerpta Medica Database), and Web of Science. The systematic search included articles published up to November 30, 2017. The following search terms were combined: “(SNP or SNPs or “single nucleotide polymorphism” or polymorphism or “genetic polymorphism” or mutation or variant or variation),” “,“ and “(IL-1β or ” interleukin-1 beta” or “IL-1 beta” or IL-1B).” Language and publication year were not restricted in our search. Finally, 1353 articles were retrieved using the aforementioned terms.2.2Eligible articles conformed to the following inclusion criteria: assessed MI as the outcome of study; assessed the association between MI and IL-1β + 3954C/T (rs1143634); presented genotype data of cases and controls with risk of MI sufficient to calculate odds ratios (ORs) and 95% confidence interval (CIs); and used a case–control design for human. Exclusion criteria included deficient genotype frequency; duplicate literature; published as a letter, comment, or review; evaluated other IL-1β SNPs and not rs1143634; case-only study; and not a human study. Two investigators separately selected the potential literature according to these criteria. When divergences appeared, the third investigator made the final decision.2.3Information from all eligible literature was extracted by 2 authors independently. The third author handled any divergences until agreement among all authors was unanimous. The following data were collected: name of first author, ethnicity of subjects, Hardy–Winberg equilibrium (HWE), sample size, genotyping method, genotype distributions in cases and controls, and the quality of study. Ethnicity was classified as Asian or Caucasian. We sent requests to corresponding authors for additional data when the primary data could not be obtained from relevant articles.2.4,25 and the Newcastle–Ottawa Scale (NOS). The adjusted criteria contained many items, such as the source of controls, the source of cases, case–control matching, sample size, genotyping method, and the HWE in controls. Two authors separately graded all included studies and any divergence was assessed by the third author. Scores ranged from 0 to 10. A study quality score ≥6 indicated “high quality,” while a study quality score <6 indicated “low quality.”The quality of eligible literature was accessed by 2 authors separately according to predetermined criteria Table , which w2.5 The control group in each included study was assessed for HWE by a Chi-square test, and a group was considered to be in Hardy–Weinberg disequilibrium at P < .05. ORs and 95% CIs were calculated to assess the strength of the association between IL-1β + 3954C/T and MI risk. The pooled ORs were used to assess allelic comparison (T vs C), a heterozygote model (TC vs CC), a homozygote model (TT vs CC), a dominant model (TT + TC vs CC), and a recessive model (TT vs TC + CC). Heterogeneity was assessed by the Q statistic  and the I2 statistic (I2 > 50% indicating a significant inconsistency). When heterogeneity existed, we carried out a random effect model (the DerSimonian and Laird method) to evaluate the pooled ORs and 95% CIs, otherwise, a fixed effect model (Mantel–Haenszel method) was performed to assess the pooled ORs and 95% CIs. Sensitivity analysis was performed by examining the effect of omitting individual studies. Begg funnel plot and Egger test were carried out to check for the publication bias (P < .05 suggested a significant bias). STATA software  was used to perform all the tests in our meta-analysis, with 2-sided P values.The meta-analysis was performed according to the PRISMA checklist and followed these guidelines.All analyses were based on previous published studies; thus, no ethical approval and patient consent are required.33.1,30 were excluded. Eventually, 9 eligible case–control publications, all conforming to the inclusion criteria, were included in our meta-analysis.A total of 1353 studies were identified from the PubMed, Cochrane, Embase, and Web of Science databases. The flow diagram in Fig. –23Table ,19 while the other studies were based on Caucasian populations.,18,20–23 The results of the HWE tests for genotypic distribution in controls are summarized in Table The 9 independent studies included in our meta-analysis included 2203 cases and 2299 controls.3.2I2 = 0%, PH = .448) and dominant model  , homozygote model , or heterozygote model  , homozygote model , and dominant model . Nevertheless, no significant association was observed in the recessive model  or heterozygote model   , but there was no evidence of a significant link in the other genetic models . No significant effects were observed in the low-quality studies  , but there was no statistically significant association in the heterozygote model, homozygote model, or recessive model . No significant association was observed in the non-RFLP subgroup . Information concerning Egger test are listed in Table We evaluated the publication bias by Begg funnel plot and Egger test. The results indicated no significant publication bias in any of the genetic models. Figure 4–23 including 2203 cases and 2299 controls, were identified and analyzed. The pooled results showed that IL-1β + 3954C/T significantly increased MI susceptibility in the allelic comparison  and the dominant model . Similar results were observed in Caucasian populations , and in subgroups from high-quality studies  and PCR-RFLP . Heterogeneity was not observed in any of the 5 genetic models used.In our meta-analysis, 9 eligible studies,In the subgroup analysis according to the quality of the studies and genotyping method, the results for the PCR-RFLP subgroup were consistent with the pooled results, and the results for the high-quality study subgroup remained similar in the allelic comparison. However, for the low-quality studies and non-RFLP subgroup, different results were observed in all genetic models. These differences may be due to the smaller sample size in these low-quality studies and the non-RFLP subgroup, which may obscure any potential association.,31 IL-1β, a crucial mediator of inflammation, is secreted by macrophages, thrombocytes, and injured endothelium,33 and plays a primary role in inflammatory reactions and atherosclerosis. Recently, many studies have demonstrated that IL-1β has pathogenic effects in many human conditions, including cardiovascular diseases.,35 Kirii et al also proved that atherosclerotic lesions in ApoE−/− IL-1β−/− mice were effectively reduced by approximately 33% compared with lesions in ApoE−/− mice, which suggests that IL-1β promots atherosclerosis. Moreover, inflammatory responses show a high interindividual difference and have been linked to single-nucleotide genetic polymorphisms in the IL-1β gene.–39 A SNP at position + 3954 of the IL-1β gene, which involves the replacement of cytosine by thymine, leads to the emergence of a low-frequency allele that is related to the overproduction of IL-1β. In addition, Pociot et al demonstrated that IL-1β + 3954C/T was correlated with elevated IL-1β expression in monocytes in vitro. Indeed, several studies have indicated a role for IL-1β + 3954C/T in risk assessments for numerous inflammatory diseases due to increased IL-1β production.,42 Thus, this polymorphism of the IL-1β gene (+3954C/T) may increase IL-1β expression, which could aggravate inflammation and finally increase the risk of MI.An increasing number of studies have found that inflammation has strong effects on the pathogenesis of atherosclerosis, CAD, and its associated complications, such as MI, which are in turn widely thought to be correlated with inflammatory processes.,21,22 However, a significant relation between IL-1β + 3954C/T and MI was found in an allelic comparison and a dominant model. Previously, Tulyakova et al and Zeybek et al and reached the similar conclusion that the T allele of IL-1β + 3954C/T was related to an increased risk of MI. IL-1β may promote atherosclerosis development through different biological functions and lead to the production of several pro-inflammatory factors such as interleukin-6, fibrinogen, and C-reactive protein. In addition, IL-1β + 3954C/T is related to increased IL-1β production.,40 All of the above studies support our meta-analysis findings that the T allele of IL-1β + 3954C/T significantly increases MI risk. Thus, individuals with the T allele of IL-1β + 3954C/T might have an increased susceptibility to MI due to a more severe inflammatory status. A subgroup analysis by ethnicity showed that a significant risk of MI susceptibility was observed in the allelic comparison, homozygote model, and dominant model among Caucasian populations. However, no significant results were observed in Asian populations. Some studies have revealed that IL-1β + 3954C/T is related to an increased risk of MI in Caucasian populations.,17 However, Daraei et al demonstrated that the TT genotype of the IL-1β + 3954C/T polymorphism was associated with a significant MI-protective effect in an Asian population, and Wang et al showed that IL-1β + 3954C/T was not correlated with MI risk in a Chinese population. MI is a multifactorial disease and polymorphisms may also have different effects on populations as a result of diverse environmental factors. Thus, ethnographic heterogeneity, along with specific hereditary backgrounds and living conditions, could determine the different effects of the IL-1β + 3954C/T polymorphism.We investigated the role of IL-1β + 3954C/T in relation to MI. No significant association between IL-1β + 3954C/T and MI was observed in a recessive model, homozygote model, or heterozygote model, which was coincident with the findings of previous studies.In our meta-analysis, we utilized a much larger total sample size than did previous studies to evaluate the effect of the IL-1β + 3954C/T polymorphism in MI. In addition, heterogeneity was not found in the pooled results. Thus, our results are more reliable than those of previous studies. However, our meta-analysis has some limitations. First, MI has a multifactorial condition and several factors were not clear in the included studies, such as smoking, living habits, and serum lipid levels. Therefore, we cannot properly assess the association between IL-1β + 3954C/T and MI in relation to these factors. Second, although we performed a systematic search to access as much of the relevant literature as possible, it is possible that we missed some studies. Finally, only 2 studies involving in Asian populations were included and they deviated from Hardy–Weinberg disequilibrium, which may have led to unreliable results for these Asian populations. Thus, more in-depth studies with large sample sizes are required to evaluate these association in Asian populations.In conclusion, our meta-analysis proved that IL-1β + 3954C/T is associated with MI susceptibility, especially among Caucasian populations.We are indebted to Huayue Lin, who helped with our study.Conceptualization: Yizhen Fang, Huabin Xie, Zhiyuan Lin.Data curation: Yizhen Fang, Huabin Xie, Zhiyuan Lin.Formal analysis: Yizhen Fang, Huabin Xie, Zhiyuan Lin.Investigation: Yizhen Fang, Huabin Xie, Zhiyuan Lin.Methodology: Yizhen Fang, Huabin Xie, Zhiyuan Lin.Software: Yizhen Fang, Huabin Xie, Zhiyuan Lin.Supervision: Huabin Xie.Validation: Yizhen Fang, Huabin Xie, Zhiyuan Lin.Writing – original draft: Yizhen Fang.


Objectives: The aim of the present study was to define the potential relationship between xeroderma pigmentosum group D (XPD) Lys751Gln polymorphisms and the risk of leukemia.Methods: A comprehensive search of Pubmed, Web of Science, EBSCO, the Cochrane Library and China National Knowledge Infrastructure was conducted to identify original articles published before March 2017 concerning the association between XPD Lys751Gln polymorphisms and leukemia risk. A literature quality assessment was performed using the Newcastle-Ottawa Scale. Heterogeneity across studies was assessed using I2 statistics. Random- or fixed-effects models were used to calculate pooled odds ratios (ORs) in the presence or absence of heterogeneity, respectively. Sensitivity analysis was used to assess the influence of individual studies on the pooled estimate. Publication bias was investigated using funnel plots and Egger’s regression test. All data analyses were performed using Stata 14.0 and Revman 5.3.Results: Fourteen studies with a total of 7525 participants  were included in this meta-analysis. We found that XPD Lys751Gln polymorphisms significantly increased the risk of developing leukemia in both dominant OR = 1.21, 95%CI [1.10–1.35], P ≤ 0.001) and heterozygote  model. An allele model showed a borderline significant increase in leukemia risk . A subgroup analysis revealed a consistent association between XPD Lys751Gln polymorphisms and leukemia risk for some genetic models in Caucasian populations, adult or chronic groups, and in almost all models of childhood or acute groups.Conclusion: Our results indicate that XPD Lys751Gln polymorphism increases the risk of leukemia, especially in childhood and acute cases. Leukemia, a common malignant disease of the hematopoietic system , can be Decreased efficiency of DNA repair is considered a crucial event in carcinogenesis . RecentlSeveral studies have reported inconsistent results on the relationship between the XPD Lys751Gln polymorphism (SNP IDs: rs13181) and leukemia susceptibility. Some studies showed a clear relationship between XPD Lys751Gln polymorphism and an increased risk of leukemia , 2016, wA computerized search of Pubmed, Web of Science, EBSCO, the Cochrane Library and China National Knowledge Infrastructure (CNKI) up to March 2017 was conducted using the following search strategy: (“XPD” or “ERCC2” or “xeroderma pigmentosum group D”), and (“polymorphism” or “variant” or “mutation”), and “Leukemia.” The search was restricted to English and Chinese language publications. A manual search of the references in the retrieved articles and relevant reviews was also conducted. A flowchart of information pertaining to identification, screening, eligibility, and the final selected datasets was constructed in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines .Studies that investigated the association between XPD Lys751Gln polymorphism and leukemia risk were included. The inclusion criteria were (1) case-control study design; (2) available genotype information of the XPD Lys751Gln polymorphism; (3) evaluation of the XPD gene polymorphism and the risk of leukemia; and (4) the distribution of genotypes among the controls was in agreement with the Hardy–Weinberg equilibrium (HWE). The major criteria for exclusion were: (1) duplication of earlier publications , (2) unpublished papers, dissertations, conference articles and reviews, (3) family-based studies of pedigrees.Data from each eligible study were extracted into Excel, and included the country of origin and ethnicity of each study population, age group (adult or childhood), subtypes of leukemia, genotyping method, numbers of cases and controls, numbers of cases and controls in the XPD Lys751Gln genotypes and results of the HWE test.The quality of the included studies was assessed by two reviewers in accordance with the Newcastle-Ottawa Scale (NOS) , which iI2 statistic. A random-effects model or fixed-effects model was used to calculate the pooled OR in the presence or absence of heterogeneity, respectively. To detect possible sources of heterogeneity and potential differences among subgroups, meta-regression and subgroup analyses were carried out with the stratification of different ethnicities, age groups and subtypes of leukemia. The significance of the pooled OR was determined through a Z-test, and p < 0.05 was considered to be statistically significant. Publication bias was investigated using funnel plots and Egger’s regression test. We also conducted sensitivity analysis to test the robustness of associations by sequentially omitting each of the included studies one at a time. All the data analysis was performed using STATA software 14.0  and Review Manager 5.3 .The combined odds ratios (ORs) and 95% confidence interval (CI) were used to evaluate the strength of the association with the risk of leukemia. Pooled ORs were performed for allelic comparison (a vs. A), dominant (aa + Aa vs. AA), recessive (aa vs. Aa + AA) and codominant (aa vs. AA and Aa vs. AA) models . Heterogeneity among the included studies was assessed using the Figure 1). A total of 14 studies  concerning XPD Lys751Gln polymorphism were included in the final evaluation (Table 1). A quality assessment of the individual studies showed that the Newcastle-Ottawa scale score ranged from 6 to 8, indicating that the quality of the methodology was generally good (Table 2).We used several search criteria to include or exclude reported studies on the relationship between XPD polymorphism and leukemia  and heterozygote models . No significant association was found in recessive  and homozygote  models. In addition, the allele model showed a borderline significant increase in leukemia risk . Moderate heterogeneity (I2: 43%–51%) was found in the no-association model group. To explore the source of this heterogeneity, a meta-regression analysis was conducted. The results revealed that the heterogeneity was not associated with ethnicity, age, clinical subtype or detection method . We further explored the source of heterogeneity by removing one study each time. The results showed that Since significant heterogeneity was identified in recessive, homozygote and alleles models, a random-effects model was used. The other genetic models were analyzed using a fixed-effects model. Overall, a significant increase in leukemia risk was identified in dominant  and dominant models . No significant association was found in recessive , homozygote  and allele models  (Table 3). In African and Asian populations, the subgroup analysis was unreliable as only two studies were available.In Caucasian populations, a significant increase in leukemia risk was found in the heterozygote model , and in almost all models of the childhmodels of the childhood group . Moderate to higher heterogeneity (I2: 46%–65%) was found in the no-association models of the adult group (Table 4).In a subgroup analysis by age group, one study was excluded as it lacked data on patients’ age . SignifiI2 = 12%, P ≤ 0.001; GlnGln + LysGln vs. LysLys: I2 = 0%, P ≤ 0.001; GlnGln vs. LysLys: I2 = 14%, P = 0.02; LysGln vs. LysLys: I2 = 0%, P = 0.003), and in some models of chronic disease  (Table 5).In a subgroup analysis by leukemia subtypes, one study was excluded as it lacked such data . SignifiThe NER pathway is a highly conserved DNA repair mechanism that removes bulky intra-strand adducts created by agents such as UV radiation and certain chemicals, including several commonly used chemotherapy agents . GeneticOur results indicate that XPD Lys751Gln polymorphism significantly increases the overall leukemia risk in dominant and heterozygote models, but not in an allele model or homozygote model. The results suggest that heterozygous mutations but not homozygote mutations of XPD (Lys/Gln) may increase genetic susceptibility to leukemia. This may be the result of a higher rate of heterozygous versus homozygous mutations (the ratio between heterozygous and homozygous mutations is 2.27 ∼ 56 in the control and 1.71 ∼ 29 in leukemia). A subgroup analysis by ethnicity showed the same result in a Caucasian population. The exact mechanism underlying the association between the susceptibility to different tumors and XPD Lys751Gln polymorphism is currently unknown. XPD is a 5′–3′ superfamily 2 DNA helicase that opens damaged DNA for bulky lesion repair in NER. The interaction of the C-terminal domain of XPD with the p44 helicase activator protein is critical for both helicase activity and the stability of the Transcription Factor II H (TFIIH) complex, which is essential for RNA polymerase II-mediated transcription initiation and the NER . The XPDCompared with previous reports , the preOur meta-analysis demonstrates that XPD Lys751Gln polymorphism significantly increases overall leukemia risk in dominant and heterozygote models, and that this polymorphism is significantly associated with almost all genetic models of childhood and acute leukemia.XH and Y-BD designed this study. XL, YC, and JS searched databases and collected the full-text papers. MW and BZ extracted and analyzed the data and wrote the manuscript. EZ, YL, XH, and Y-BD reviewed the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.