Purpose: To provide a comprehensive account of the association of five Lymphotoxin-α (LTA) gene polymorphisms  with susceptibility to cancer.Methods: A literature search for eligible candidate gene studies published before 28 February 2020 was conducted in the PubMed, Medline, Google Scholar and Web of Science. The following combinations of main keywords were used:  AND (polymorphism OR mutation OR variation OR SNP OR genotype) AND . Potential sources of heterogeneity were sought out via subgroup and sensitivity analysis, and publication bias were estimated.Results: Overall, a total of 24 articles with 24577 cases and 33351 controls for five polymorphisms of LTA gene were enrolled. We identified that rs2239704 was associated with a reduced risk of cancer. While for other polymorphisms, the results showed no significant association with cancer risk. In the stratified analysis of rs1041981, we found that Asians might have less susceptibility to cancer. At the same time, we found that rs2239704 was negatively correlated with non-Hodgkin lymphoma (NHL). While, for rs909253, an increased risk of cancer for Caucasians and HCC susceptibility were uncovered in the stratified analysis of by ethnicity and cancer type.Conclusion:LTA rs2239704 polymorphism is inversely associated with the risk of cancer. LTA rs1041981 polymorphism is negatively associated with cancer risk in Asia. While, LTA rs909253 polymorphism is a risk factor for HCC in Caucasian population. LTA gene is located on the short arm of chromosome 6 (6p21.3) [LTA gene at the 804 (C/A) position of exon 3 in codon 26, causing the amino acid threonine to be asparagine, which may be related to the transcriptional regulation of LTA, then activate the lymphocytes and induce apoptosis [LTA gene at 252 (A/G) position in intron 1, which may lead to increase in the transcriptional activities of LTA [LTA expression, which may affect subsequent inflammatory responses and immunomodulatory diseases, including cancers [Increasing studies have demonstrated that a number of proinflammatory cytokines could be associated with the development of cancer ,2. Lymph6p21.3) . The prep21.3 . T(6p21.3) . SNP rs1 cancers ,10.LTA polymorphisms and cancer [LTA rs1041981, rs2239704 and rs2229094 polymorphisms were associated with the increased risk of cancers [LTA polymorphisms and cancer [LTA gene and cancer susceptibility.There are ample evidences that have demonstrated the association between d cancer . However cancers . Howeverd cancer ,28,31,32LTA polymorphisms and the risk of cancer (up to 28 February 2020) with the following keywords:  AND (polymorphism OR mutation OR variation OR SNP OR genotype) AND . The language of enrolled studies was restricted to English. After carefully screening, five polymorphisms were left for further investigation.We conducted a systematic literature search on PubMed, Medline, Embase, Google Scholar and Web of Science to retrieve all eligible publications on the association between LTA polymorphisms and cancer risk; (2) publications focusing on population genetic polymorphisms; (3) articles with sufficient genotype data to assess odds ratios (ORs) and the corresponding 95% CIs; (4) blood sample only for SNP analysis; (5) the control subjects satisfied HWE. The major exclusion criteria were: (1) case-only studies, case reports or reviews; (2) studies without raw data for the LTA genotype.Articles enrolled in our meta-analysis satisfied the following inclusion criteria: (1) case–control studies that evaluated the association between LTA genotypes. Ethnicity was categorized as ‘Asian’, ‘Caucasian’, and ‘Mixed’.Two investigators (Jingdong Li and Yaxuan Wang) independently extracted the data from each study. All the case–control studies satisfied the inclusion criteria and consensus for any controversy was achieved. The data from the eligible articles comprise the first author’s name, year of publication, ethnicity, source of control, cancer type and numbers of cases and controls in LTA polymorphisms and cancer was evaluated using summary ORs and the corresponding 95% CIs in allelic (B vs. A), dominant (BA + BB vs. AA), and recessive (BB vs. BA + AA) models . The Cochrane’s Q-statistic test was used to assess the heterogeneity between studies, and the inconsistency was quantified with the 2I statistic. The substantial heterogeneity was considered significant when 2I > 50% or PQ ≤ 0.1, then, a random-effects model was used; otherwise, the fixed-effects model was applied. Subgroup meta-analysis were performed by cancer type, ethnicity, genotyping, HWE and the source of control. We also conducted sensitivity analysis to assess stability of the results by omitting one study each time to exclude studies. HWE was estimated by the asymptotic test, and deviation was considered when P<0.05. The potential publication bias of the eligible studies was evaluated by Begg’s and Egger’s regression test quantitatively. Trial sequential analysis (TSA) was performed as described by Xie et al. [P<0.05 was considered statistically significant.The risk between the e et al. . The reqLTA gene , a total of 24 articles (including 43 case–control studies) with 24577 cases and 33351 controls met the inclusion criteria [LTA polymorphisms included in our meta-analysis. Newcastle–Ottawa scale (NOS) was used to evaluate the quality of the enrolled studies, as shown in The study selection processes were presented in criteria . Sixteencriteria ,32. Tablrs1041981P=0.027, P=0.002). Moreover, when the subgroup analysis was performed based on source of controls, hospital-based control group was significantly related to a decreased risk of cancer in allelic contrast  and dominant model  (The pooled results based on six included studies  (includiP=0.000) .The pooled results based on four included studies ,17–19 (iP=0.003), dominant model  and recessive model . Furthermore, in the stratification analysis by cancer type, we observed that rs2239704 reduced the risk of NHL in allelic contrast  and recessive model . Moreover, when the subgroup analysis was performed based on ethnicity, source of control and genotyping, we found mixed ethnicity was significantly related to a reduced risk of cancer in allelic contrast , dominant model  and recessive model  (The pooled results based on eight included studies  (includiP=0.013) .The pooled results based on four included studies ,25–27  and recessive model . In addition, in the stratification analysis by ethnicity, we observed that rs909253 polymorphism was significantly related to an increased risk of Caucasian ethnicity in allelic contrast , and mixed ethnicity in allelic contrast , dominant model . Moreover, in the stratification analysis by source of control, genotyping and HWE, null result was found . In addition, Begg’s test and Egger’s regression test were performed to evaluate the publication bias. As for rs1041981, rs2229094, rs2239704, rs746868 and rs909253, no evidence of publication bias was identified (Supplementary Table S3).To evaluate random errors, we performed TSA . This anLTA gene polymorphisms  and the risk of cancer. We identified that rs2239704 was inversely associated with the risk of cancer under different genetic models. However, for LTA rs1041981, rs2229094, rs746868, rs909253 polymorphisms, no significant association with cancer risk was uncovered.In the present study, a total of 24 articles including 43 case–control studies were enrolled to validate the association between five In subgroup meta-analysis stratified by cancer type, we found that rs2239704 was significantly reduced NHL susceptibility. Huang et al. reported rs2239704 polymorphism was correlated with cancer and positive association in North Americans . HoweverIn the stratified analysis of rs1041981, we found that Asians might have less susceptibility to cancer. Unlike the study by Huang et al. , we exclLTA rs2229094 and rs746848, only four studies reported their relationship with cancer in each group. No significant results were found. Huang et al. reported positive association between rs2229094 and cancer risk [For cer risk , which ccer risk . BecauseLTA rs909253 and the linkage disequilibrium with other SNPs.Although the overall analysis of rs909253 indicated a null result for cancer risk, the risk of cancer for Caucasians and HCC susceptibility were significantly increased in the stratified analysis by ethnicity and cancer types. In addition, some of the control groups did not match HWE, we can not exclude the possibility that may cause the bias. Then, subgroup analysis by HWE showed that HWE status did not cause the bias of results. Huang et al. did not report the relationship of rs909253 and cancer risk, because it might be present in high linkage disequilibrium with other four SNPs ,9. HowevLTA polymorphisms with various cancers, and we could not get enough data for some cancer types. Fourth, we did not assess the linkage disequilibrium, which might not reflect the real function correctly. In future, more well-designed case–control studies are needed to investigate the functions of LTA polymorphisms.In the present study, we have put great effort on carefully searching for eligible studies. In order to obtain more accurate and reliable results, we conducted a comprehensive search to verify more eligible studies. Then, we used NOS to evaluate the quality of the included studies, eliminate low-quality studies and improve overall research quality. In order to provide the sources of heterogeneity, subgroup analysis was performed by ethnicity, cancer type, source of controls, genotyping and so on. In addition, sensitivity analysis was used to confirm the stability of the studies. Egger’s and Begg’s tests were used to assess publication bias. However, several limitations in our study should be noted. First, small sample size limits the reliability of the results for some polymorphisms. Second, we just included the studies published in English, which may influence the effects of the polymorphisms. Third, we mainly evaluated the relationship between LTA rs2239704 polymorphism is inversely associated with the risk of cancer, as is LTA rs1041981 polymorphism in Asia. While, LTA rs909253 polymorphism is a risk factor for HCC in Caucasians. Further studies with larger sample size are needed to confirm these findings.Our meta-analysis suggests that Click here for additional data file.


A nonlinear relationship was suggested between SBP and SCD. The results persisted in most subgroup and sensitivity analyses. There was no evidence of publication bias. This meta-analysis found an increased risk of SCD with hypertension diagnosis and increasing SBP. Future studies should clarify the association for DBP and the shape of the dose–response relationship between blood pressure and SCD.Cardiovascular disease is the leading cause of death worldwide, while sudden cardiac death (SCD) accounts for over 60% of all cardiovascular deaths. Elevated blood pressure and hypertension have been associated with increased risk of SCD, but the findings have not been consistent. To clarify whether blood pressure or hypertension is associated with increased risk of SCD and to quantify the size and the shape of any association observed. PubMed and Embase databases were searched for published prospective studies on blood pressure or hypertension and SCD up to 30 April 2018. Summary relative risks (RRs) and 95% confidence intervals (CIs) were calculated using a random effects model. The meta-analysis included 2939 SCDs among 418,235 participants from 18 studies. The summary RRs were 2.10  contains supplementary material, which is available to authorized users. Cardiovascular disease (CVD) is a major cause of morbidity and mortality worldwide , 2. SuddElevated blood pressure is the single largest contributor to the global burden of disease, accounting for two-thirds of strokes, half of coronary heart disease (CHD) cases and a total of 9.4 million global deaths per year –10. SeveThe study was conducted according to the preferred reporting items for systematic reviews and meta-analyses (PRISMA) statement . The proPubMed and Embase were searched from inception up to 30 April 2018. The search terms used are shown in Supplementary Table 1. The reference lists of the included publications were also screened with additional studies being identified.Prospective cohort studies, retrospective cohort studies, nested case–control studies and case-cohort studies of general adult populations that provide risk estimates for the association between blood pressure or hypertension and the risk of SCD with adjustment for at least one confounding factor were included. When multiple articles had been published on the same exposure from the same study, the one with the largest number of participants and/or cases or adjusted for SCD risk factors with higher level of evidence was chosen.The following data were extracted from each included study: name of the first author, study publication year, country where the study was conducted, name of the study, study period and the length of follow-up, sample size and participant characteristics , number of cases, type of exposure  or diastolic blood pressure (DBP)), subgroup(s), RRs and 95% confidence intervals (CIs) and confounder(s) adjusted for in the analysis. For studies in which RRs and 95% CIs were not readily available, conversions were made using listed regression coefficients, standard errors, p values and/or t/z values.Study quality was evaluated using the Newcastle–Ottawa scale . ‘Demonsp value < 0.05 was considered statistically significant. The method by Greenland and Longnecker [Random effects models  which acngnecker  was adopngnecker  were usengnecker . Nonlinengnecker .2 statistic [2 values of approximately 25%, 50% and 75% correspond to low, moderate and high degrees of heterogeneity, respectively. Subgroup and sensitivity analyses were conducted to evaluate the result consistency across various study characteristics and to investigate sources of heterogeneity between studies. Pre-specified factors included: sex, publication year, length of follow-up, geographic location, number of cases, study quality, baseline CHD status, hypertension definition, SCD definition, age and other adjusted confounders. Meta-regression analyses were used to test for differences between subgroup analyses and to examine the impact of individual characteristics on the overall effect size and/or their contributions to the detected heterogeneity [p value < 0.10 was considered as an indication of potential publication bias. All statistical analyses were completed using the Stata software, version 13.1 .Heterogeneity between studies was assessed using the Q statistic and the Itatistic . I2 valuogeneity . Sensitiogeneity  and by iA total of 5691 records were screened, of which 5689 records were identified from database search and two additional studies , 27 wereDetailed study-level characteristics and the extracted data are presented in Supplementary Table 2 and Supplementary Table 3 for studies on hypertension and blood pressure, respectively. One study was a nested case–control study  and all 2 = 56.7%, pheterogeneity = 0.02). There was neither evidence of publication bias with Begg’s test (p = 0.47) or Egger’s test (p = 0.14), nor asymmetry indicated by the funnel plot . When individual studies were excluded from the analysis one at a time, the summary RRs ranged from 1.96 (95% CI 1.63–2.36) when the study by Karppi et al. [Ten studies (nine publications) –31, 33 ii et al.  was exclThe meta-analysis of blood pressure and the risk of SCD was comprised of 11 studies (seven publications) , 30, 32.2 = 45.5%, pheterogeneity = 0.07) per 20 mmHg increment of SBP. Begg’s test (p = 0.92), Egger’s test (p = 0.50) and the inspection of the funnel plot  suggested no evidence of publication bias. The influence analysis presented a range of summary RRs from 1.24 (95% CI 1.17–1.32) when the study by Bogle et al. [nonlinearity < 0.001), with a slightly steeper increase in risk from around 140 mmHg  , 30, 32 e et al.  was excl2 = 83.4%, pheterogeneity = 0.002). No evidence of publication bias was indicated . There was no evidence of a nonlinear association between DBP and SCD  of 7.7 (8) out of 9 across all included studies (Supplementary Table 6).heterogeneity = 0.03). No significant heterogeneity was present between the remaining subgroups. Nevertheless, most analyses showed substantial heterogeneity within each subgroup. One exception was male-only studies, which had zero heterogeneity across all types of exposure.The subgroup and sensitivity analyses results are shown in Table 2 = 88.0%, pheterogeneity < 0.001) for prevalent hypertension and 1.25  and 1.08  per 20 mmHg and 10 mmHg increments in SBP and DBP, respectively. There was no change to the direction or the significance of the associations.A sensitivity analysis was conducted to incorporate one study with a combined outcome of SCA/SCD , includiThis is, to the best of our knowledge, the first systematic review and meta-analysis of cohort studies on blood pressure, hypertension and the risk of SCD. The results suggested a twofold increase in risk for SCD with prevalent hypertension and a 28% increase in risk for SCD per 20 mmHg increment in SBP, while no significant association was established between DBP and SCD. The results were robust regarding the association between hypertension/SBP and SCD, whereas the analysis for DBP and SCD was limited by the few studies included. Additionally, there was evidence of a nonlinear relationship between SBP, but not DBP, and SCD, with a slightly steeper association at higher levels. Although the nonlinear analysis contained only two small male studies, there was a clear dose–response relationship. Overall, the analyses involved good-to-high-quality studies from three continents with close to 3000 SCDs among a total of more than 410,000 participants. There was no evidence of publication bias; and the relationships were also maintained in most subgroup and sensitivity analyses.The findings are partly consistent with a pooled analysis of 61 prospective studies  , which fThe risk estimates differed substantially when the studies were stratified by adjustments for resting heart rate and AF, which is consistent with previously published systematic reviews and meta-analyses that established positive relationships between resting heart rate/AF and SCD , 36. NonSeveral potential mechanisms could explain an association between high blood pressure or hypertension and increased risk of SCD. First, chronic elevated blood pressure induces adaptive myocardial structural changes, which in turn leads to a cascade of anatomical and functional changes causing left ventricular hypertrophy (LVH) . LVH is Limitations of the current meta-analysis include potential confounding, exposure and/or outcome misclassification and between-study heterogeneity. Although only studies with adjusted RRs were included and in-depth analyses suggested little heterogeneity among subgroups, residual confounding cannot be entirely ruled out. Some studies may have also inappropriately adjusted for intermediate factors on the causal pathway from elevated blood pressure to SCD, such as LVH, CHD, HF or AF.Most studies defined hypertension based on blood pressure measurements and antihypertensive use. Medication use was mainly self-reported, which may have under-estimated the actual number of people with hypertension and the RRs. A few studies also adopted higher blood pressure cut-off values for hypertension diagnosis. However, the sensitivity analysis suggested no significant heterogeneity among studies with different hypertension definitions. Moreover, SCD case adjudication has always been challenging due to often unwitnessed deaths and/or inadequate information . The incThere was considerable between-study heterogeneity as well. Nevertheless, except for the analyses on DBP which only included three studies, the reported heterogeneity was mainly due to differences in the size of the association rather than the overall direction.This systematic review and meta-analysis has a number of strengths. The inclusion of cohort studies ensured that the exposure preceded the outcome, avoided recall bias and reduced the potential for selection bias. The meta-analysis included close to 3000 SCDs in over 410,000 participants, providing sufficient statistical power to detect even moderate associations between hypertension/SBP and SCD. The high quality of the included studies and the detailed subgroup and sensitivity analyses further ascertained the robustness of the results. Finally, the dose–response analysis provided additional evidence for a dose–response relationship between elevated blood pressure and the risk of SCD.Further studies should aim to clarify the association between DBP and SCD and the shape of the dose–response relationship between blood pressure and SCD. As most included studies were from Europe and North America, additional studies are needed to clarify these associations in other geographic locations and among different ethnicities. Only studies from the general population were included in this systematic review and meta-analysis. Further studies could clarify whether modifying factors such as the use of drug therapies and its related biochemical and/or physiological influences would affect the association between hypertension or blood pressure and SCD. Given that hypertension is linked to various lifestyle factors, future studies should also address the association between lifestyle components and SCD directly, as it is possible that elevated blood pressure might be a mediator of the adverse effects of other risk factors. Public health policies and interventions to further address the hypertension epidemic are recommended to reduce the burden of SCD.Supplementary material 1 (DOCX 696 kb)Below is the link to the electronic supplementary material.


The aims of this meta-analysis are (i) to estimate the pooled relative risk (RR) of developing non-affective psychotic disorder (NAPD) and affective psychotic disorder (APD) among migrants and their children; (ii) to adjust these results for socioeconomic status (SES); (iii) to examine the sources of heterogeneity that underlie the risk of NAPD.We included population-based incidence studies that reported an age-adjusted RR with 95% confidence interval (CI) published 1 January 1977–12 October 2017 and used a random-effects model.n = 43), Israel (n = 3), Canada (n = 2) and Australia (n = 1). The meta-analysis yielded a RR, adjusted for age and sex, of 2.13 (95% CI 1.99–2.27) for NAPD and 2.94 (95% CI 2.28–3.79) for APD. The RRs diminished, but persisted after adjustment for SES. With reference to NAPD: a personal or parental history of migration to Europe from countries outside Europe was associated with a higher RR  than migration within Europe . The corresponding RR was lower in Israel  and Canada . The RR was highest among individuals with a black skin colour . The evidence of a difference in risk between first and second generation was insufficient.We retrieved studies performed in Europe (Positive selection may explain the low risk in Canada, while the change from exclusion to inclusion may do the same in Israel. Given the high risks among migrants from developing countries in Europe, social exclusion may have a pathogenic role. It is not surprising, therefore, that a recent meta-analysis found migrants to be at increased risk of developing mood disorders  , and that adjusts for socioeconomic status (SES) in the country of destination. Two competing theories have been proposed to explain the low SES of individuals who develop a psychotic disorder (PD): social causation (stress) and social selection . Since there is little evidence of an association between parental SES and risk of psychosis, the mechanism of social selection may be more important than that of social causation  of developing APD, NAPD, or any of these disorders, among international migrants and their children; (ii) to adjust these results for SES; and (iii) to examine sources of heterogeneity with reference to the RR of NAPD, such as generational status (first or second), region of destination, region of origin, developmental level of country of origin, skin colour, refugee status and sex.A previous meta-analysis reported that the risk of NAPD did not differ significantly between first- and second-generation migrants  to report a risk ratio  with 95% confidence interval (CI) for the incidence or prevalence of APD and/or NAPD among migrants in a circumscribed geographical area ; (ii) to adjust the risk ratio for differences in age between migrants and the reference population (or to provide data that make this adjustment possible); and (iii) to have been published in a peer-reviewed journal. For the purpose of the present study, we used incidence studies only.For details of the study selection, see online Supplementary Methods and Supplementary Fig. S1.Two authors (FT and EV) evaluated the quality of the articles independently, using the criteria given in online Supplementary Table S1 (range: 0–15). The inter-rater reliability of this procedure was good (intraclass correlation coefficient = 0.85). The averages of these scores were divided into tertiles and considered as indicating low (⩽9), medium (>9 to ⩽11) or high (>11) quality.Two authors (FT and JPS) extracted the effect sizes independently. They recorded information about diagnosis , country of destination, country or region of origin, developmental level of country of origin, skin colour, refugee status, sex, birthplace (first or second generation), adjustment for age, sex and/or SES. In case of any discrepancy, consensus was reached by discussion.First, we calculated the pooled RR for the development of any PD , NAPD and APD among migrants and their children worldwide, adjusted for age and sex .The analyses were performed using the ‘metan’ procedure of STATA . Funnel plots were used to investigate possible publication bias .et al., et al., Since only two studies adjusted the results for SES at birth  migrants. Average effect estimates were calculated for migrants and their children according to region of destination  and region of origin . Owing to the small number of studies presenting effect sizes for migrants from different regions, a comparison of effect sizes in the same study (as for SES) was not feasible here. In analysis 5.4, we compared the effect of migration to Europe from countries outside Europe to that of  migration within Europe.We classified the developmental level of the country or region of origin according to the definitions used by the United Nations Conference on Trade and Development  white, i.e. from Europe, North America or Australia; (2) white other ; (3) black (i.e. from the Caribbean or sub-Saharan Africa); (4) other  or mixed ; and (5) unknown . For the comparison of groups with different skin colours, we selected studies with at least three of these groups.Analysis 5.7 investigated the effect of refugee status, and analysis 5.8 that of sex (migrants and their children). Lastly, the results of analyses 5.5  and 5.6 (skin colour) were adjusted for SES in analyses 6.1 and 6.2.n = 37), Israel (n = 3), Canada (n = 2) or Australia (n = 1). Forty-seven articles concerned the risk of NAPD and four the risk of APD. At least 20 633 first- and second-generation migrants developed a PD (affective or non-affective) during 35 890 528 person-years at risk. At least 96 850 members of the reference population did so during 623 587 721 person-years. From six studies, the relevant numbers of cases and/or denominators could not be derived .p = 0.01). We therefore performed a sensitivity analysis by excluding studies with standard errors >0.2 and by omitting a study that reported an excessive RR, >10.0 . When we also excluded the low-quality studies, the RR of any PD was 1.97 (95% CI 1.84–2.10). Using robumeta, this figure was 2.23 (95% CI 1.80–2.76).The funnel plot of studies of the risk of any PD showed evidence of publication bias . Using robumeta, the RR was 2.18 (1.86–2.55). When we also excluded the low-quality studies, the RR of NAPD was 1.98 (1.86–2.12), with robumeta: 2.19 (1.82–2.64).Application of these analyses to studies of the risk of NAPD yielded similar results: when we excluded the studies with standard errors >2.0 and the Harrison Eight studies adjusted findings for SES. Although adjustment for SES resulted in a decrease in the RR of any PD , this measure remained significant. The same was true when the RR of APD and NAPD was adjusted for SES .When we considered all the available effect sizes with reference to NAPD, the RR of psychosis among first-generation migrants was significantly higher than that among second-generation migrants. However, when we restricted the analyses to those publications that reported effect sizes for both generations, the difference in RR became smaller and statistically not significant .Table 3and their children, except one analysis on refugees.The results presented below apply to migrants The RR of developing NAPD was lowest in Israel and Canada and significantly higher in all European destinations . Region p = 0.24).In Europe, the RR was higher for migrants and their children who came from countries outside Europe  than among their peers from within Europe . The funnel plot for the first analysis (from outside Europe to Europe) showed no evidence of publication bias  (online Supplementary Table S2).Owing to the small numbers of studies, it was not possible to repeat all the analyses using ‘robumeta’. However, where this was possible, the conclusions remained the same.Although findings indicate that migrants are at increased risk of developing APD and NAPD, the findings were heterogeneous, which precludes a conclusion about ‘the’ RR among migrants in general. While there was no evidence of a greatly increased risk among migrants in Israel or Canada, the risk among migrants in Europe, in particular those from developing countries outside Europe and/or those with a black skin, was greatly increased. Overall, the risk became somewhat attenuated after adjustment for SES at first contact. A comparison of risks for first- and second-generation migrants based on effect estimates derived from the same study yielded no significant difference. This indicates that membership of a disadvantaged ethnic minority group, rather than a personal history of migration, is an important determinant of risk. There was insufficient evidence to conclude that there is a difference in risk between refugees and non-refugees.et al., et al., It is unlikely that the findings can be explained by bias. Some researchers , it was often possible to avoid confounding by study.Our meta-analysis has several limitations. First, our results apply mainly to Europe. Of note, studies of first hospital admissions published before 1977 reported a modestly increased risk of schizophrenia among migrants to the USA. Malzberg , for exaet al., et al. , we do not suggest that this group is not exposed to prejudice or discrimination.et al., et al., et al., How should the findings be interpreted? It was previously thought that individuals with a genetic predisposition for psychosis are more likely to migrate than others  social disadvantage .We conclude that the increased psychosis risk among migrants and their children is a major public health problem. Since the lifetime morbid risk for psychosis (affective or non-affective) is about 1–2% for Europeans, our findings suggest that this risk may be 3–6% for migrants from outside Europe. The challenge is to advance our understanding of the underlying mechanisms and to find ways for prevention.


Although numerous replication case-control studies have attempted to determine the association between Factor V Leiden (FVL) 1691G > A mutation and susceptibility to Recurrent pregnancy loss (RPL), there have been confliction among the results of various ethnic groups. To address this limitation, here we implemented first meta-analysis to provide with consistent conclusion of the association between FVL 1691G > A mutation and RPL risk.After a systematic literature search, pooled odds ratio (OR) and their corresponding 95% confidence interval (CI) were used to evaluate the strength of the association. Additionally, meta-regression analyses were performed to find potential source of heterogeneity.P < 0.001), over-dominant , allelic , and heterozygote  models. Moreover, a significant association of dominant , over-dominant , and heterozygote  models was found in the Iranian population. The subgroup analysis indicated strong significant association in Asian, European, Africa population, and case-control studies but not in South Americans and cohort studies.In this meta-analysis, 62 studies, containing 10,410 cases and 9406 controls, were included in quantitative analysis. Overall population analysis revealed a significant positive association in the dominant (OR = 2.15, 95% CI = 1.84–2.50, The FVL 1691G > A mutation and the risk of RPL confers a genetic contributing factor in increasing the risk of RPL, particularly in Iranians, except for South Americans. Recurrent pregnancy loss (RPL) is a heterogeneous disorder which affects women of reproductive age. Recently, The American Society of Reproductive Medicine has defined RPL as two or more than two failed pregnancies before the 20th week of pregnancy –3. OveraAlthough pregnancy as a physiological condition is associated with a hypercoagulable state, and the contact between placenta and maternal circulation is crucial for the establishment of a successful pregnancy, but any abnormality in this circulation, especially abnormal blood clotting in the small placental blood vessels, may results in RPL , 10. DurFVL mutation shows an autosomal dominant pattern which occurs by substitution of guanine by adenine (CGA--- > CAA) at the nucleotide 1691 in the exon 10. As a result of this missense mutation, arginine (Arg) at amino acid 506 is substituted with glutamine (Gln), leading to generation of FVL resistant to the activated protein C (APC). APC is a natural anticoagulant which in normal situation cleaves activated factor V at amino acid 506 and makes it inactive –20.Studies have shown that FVL mutation increases the risk of venous thrombosis 7 times in heterozygote and 80 times in homozygote carriers. In addition, it has been reported that this mutation increases the risk of pre-eclampsia in FVL carriers , 22. TheIn spite of all findings, still the exact association between FVL mutation and the risk RPL is unclear and several investigators worldwide try to clarify this question. Therefore, here we conducted the first and the most comprehensive meta-analysis on the association between FVL 1691G > A mutation and risk of RPL by exerting 62 studies encompassing 10,410 cases and 9406 health control to achieve more reliable conclusion.Ethical approval is not necessary for this meta-analysis. The current meta-analysis was conducted according to the Preferred Reporting Items for Systematic reviews and Meta-Analyses (PRISMA) statement , includiA comprehensive systematic search in the ISI Web of Science, Scopus, and PubMed/Medline databases was conducted to retrieve all publications evaluating the associations between FVL 1691G > A mutation and susceptibility to RPL prior to May 2020. The following combinations of key words were used:  AND (“Factor V Leiden” OR “FV Leiden” OR “1691G > A” OR “rs6025”) AND . In spite of detailed search, a manual cross-check of eligible studies and reviews was carried out to include other potential studies. Original data in English language and human population studies were collected.Primary search strategy generates 1266 studies that were exported into Endnote X8 software. The duplicated studies were removed and title & abstract of remaining studies were reviewed by two investigators and irrelevant studies were excluded. Full-text verification was performed if we could not classify studies based on title & abstract. Any disagreements during study selection were discussed and resolved by consensus.Studies considered eligible if they met the following inclusion criteria: a) Studies concerning the association between FVL 1691G > A mutation and susceptibility to recurrent pregnancy loss as the main outcome; b) Studies that their case group have recurrent pregnancy loss (two or more times of abortion); c) Studies with case-control and cohort design; d) Studies reporting sufficient data of genotype or allele frequency that could confer feasibility of calculating the odds ratios (ORs) and 95% confidence intervals (CIs). On the other hand, duplicates, case reports, book chapters, reviews, letter to editor, studies with insufficient data, and abstracts were all excluded.According to a standardized extraction form, the following data were independently extracted by two investigators: the first author’s last name, journal and year of publication, country of origin, ethnicity, allele and genotype frequency in cases and controls, mean or range of age, genotyping method, and total sample size of cases and controls. The third investigator finalized the extracted data, and potential discrepancies were resolved by consensus. For quality assessment of the included publications, the Newcastle-Ottawa Scale (NOS) was applied . In thisP value< 0.10 was considered as statistically significant) [2) tests was used. The fixed-effected model (FEM) was used if PQ-statistic > 0.10 or I2 was< 50%; otherwise, the random-effected model (REM) was applied. In order to assessed the predefined sources of heterogeneity among included studies, subgroup analysis and meta-regression analysis based on year of population, the continent of the study population, and genotyping method were performed. Additionally, sensitivity analysis was conducted in presence of heterogeneity [P value< 0.05 was considered as statistically significant) [Deviation from Hardy–Weinberg equilibrium (HWE) for distribution of the genotype frequencies was analyzed by χ2-test in the control group. The strength of the association between FVL 1691G > A mutation and RPL risk was evaluated by the pooled OR and its corresponding 95% CI. Different comparison models for FVL 1691G > A mutation were as follow: dominant model (AA+GA vs. GG), over-dominant model (GA vs. GG + AA), allelic model (A vs. G), and heterozygote (GA vs. GG). It should be noted that due to the AA genotype frequency of zero in both cases and controls, the recessive and homozygote models were not calculable. Presence of heterogeneity between included studies was estimated by Cochran’s Q-statistic (ificant) . Besidesogeneity , 29. Pubificant) , 31. The2 qualified studies were included in the quantitative analysis, of which two studies were detected by cross-check of eligible studies and reviews [60 studies had case-control design. Tables The four-phase search and screening process of the literatures based on the PRISMA statement is depicted in the Fig.  reviews , 33. All2 studies with 10,410 cases and 9406 controls included in quantitative analysis of the association between FVL 1691G > A mutation and the risk of RPL. Of those, 25 studies were in Asian countries [6 studies were conducted in European countries [5, 95% CI = 1.84–2.50,P < 0.001, FEM), over-dominant model , allelic model , and heterozygote model  , over-dominant model , and heterozygote model  but not allelic model   were in the first rank with respect to sample size and the number of studies, therefore we performed separate analysis. Our results found a significant association between FVL 1691G > A mutation and increased risk of RPL in this population under dominant model , South America (6 studies), Africa (4 studies) and Oceania (1 article). Since there was only one study for Oceania, we exclude it from the subgroup analysis. The final results revealed strong significant association between FVL 1691G > A mutation and the risk of RPL in Asian, European, and Africa population, but not in South Americans , Europe , over-dominant model , allelic model , and heterozygote model . However, no significant association was observed in cohort studies  between various genes and variations . HoweverThe FVL 1691G > A mutation is a G-to-A point mutation at nucleotide 1691 in the factor V gene, that results in the single amino-acid replacement Arg506Gln, leading to resistance to be cleaved and, therefore, inactivation by APC and promoted susceptibility to clotting , 90. ThiP = 0.38), among others. On the other hand, a comprehensive systematic review and meta-analysis in 2016 [In this meta-analysis, 62 studies, containing 10,410 cases and 9406 controls, were included in quantitative analysis. The analysis of overall population indicated that all genetic comparisons of the FVL 1691G > A mutation, including dominant model (OR = 2.15), over-dominant model (OR = 1.88), allelic model (OR = 2.05), and heterozygote model (OR = 1.97) significantly increased the risk of RPL susceptibility. In 2015, Sergi et al.  in 2019  in 2016 , by exerOn the other side, the analysis was also performed in the Iranian population, containing 9 publications with 1409 cases and 1160 controls. The previous meta-analysis in Iranian population by Kamali et al.  in 2018,There was a degree of heterogeneity during the overall analysis. From statistical perspective, this heterogeneity describes the variability between included studies and may originate from clinical or methodological heterogeneity, from other unreported, unknown study characteristics, or may be due to chance. Therefore, for finding any sources of heterogeneity and attenuating their effects, we conducted subgroup analysis and weighted meta-regression. Collectively, the results of meta-regression showed that none of the parameters, including publication year, the continent of the study population, and genotyping methods were the expected source of heterogeneity. However, subgroup analysis reduced heterogeneity in all groups and explained part of the observed heterogeneity expect Asians and studies with cohort design. Furthermore, the other way of dealing with statistical heterogeneity, which we used in our analysis, was to incorporate “Random” term to account for it in a random-effects. Random effect model typically produces more conservative estimates of the significance of a result . As it gives proportionately higher weights to smaller studies and lower weights to larger studies than fixed effect analysis.To address the limitations in the current meta-analysis, it should be stated that, first our literature search was limited to only studies published in English language. Second, there was a degree of heterogeneity during the overall analysis. But not in all subgroup analyses, indicating the role of genetic diversity and other confounders in susceptibility to RPL. Third, as this meta-analysis a crude estimation of the association between FVL 1691G > A mutation and the risk of RPL, thus the roles of age, paternal genetic impression, environmental factors, and the effect of gene-gene interactions in conferring the susceptibility risk to RPL were neglected.Considering all the facts, this meta-analysis, the first one of its type to our best knowledge, retrieved 62 studies, encompassing 10,410 cases and 9406 health controls, to find a consistent result of the association between FVL 1691G > A mutation and risk of RPL. Our results indicated statistically significant increased risk of RPL in the overall analysis. The increased susceptibility to RPL was also observed in Iranian, Asian, European, Africa populations, and studies with case-control design, but not in South Americans and studies with cohort design. Further experiments, alongside with inclusion of additional studies with large sample sizes, should consider the role cofounders in susceptibility to RPL.


Background: The causal relationship between serum lipid levels and the risk of Parkinson's disease (PD) remains largely uncertain. We summarized the existing controversial evidence on this topic.Methods: We searched the electronic databases for observational studies from January 1988 to March 2020. We applied random-effects models to calculate pooled relative risk (RR) with their 95% confidence intervals (CI). Random-effects dose-response meta-analyses were further conducted to explore the dose-risk relationship.Results: Twelve cohort studies and three case-control studies were included in this meta-analysis. Higher levels of serum low-density lipoprotein cholesterol (LDL-C) were inversely associated with the subsequent risk of PD , whereas, there were no associations between serum levels of total cholesterol (TC) , high-density lipoprotein cholesterol (HDL-C) , or triglycerides (TG)  and the risk of PD. Further dose-response meta-analysis revealed that every 38.6 mg/dL (1mmol/L) increase in serum LDL-C correlates with a 7% decreased risk of PD.Conclusions: Our paper supports the protective effect of higher serum LDL-C on the subsequent risk of PD. More prospective cohort studies are warranted to confirm the conclusion, and further fundamental researches are needed to elucidate the underlying biological mechanisms. Parkinson's disease (PD) is a severe neurodegenerative disease, affecting 1% of the population above the age of 60 , the majClinically, disrupted lipid metabolism can lead to abnormal serum lipid levels of TC, LDL-C, HDL-C, and TG. However, the causal relationship between serum lipid levels and the risk of PD has not yet been determined. Since the first attempt in 1994 , many epAlthough a meta-analysis in 2013 observed no association between serum cholesterol and the risk of PD , more riOur meta-analysis followed the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) statement see  20). Th. Th20). Studies were included if they fulfilled the following criteria:  cohort oTwo authors (ZJ and XRX) independently extracted the following data from included studies: first author and publication year; study location; study design; cohort name; serum lipid parameters; data type; follow-up time; number of participants/person-years, number of cases; age; outcome definition (relative risk[RR]/odds ratio[OR]/hazard ratio[HR]); covariates adjusted; levels of exposure, along with corresponding risk estimates and 95% CI for each category. Disagreements were settled by mutual discussion. If articles selected in this paper report concentrations of serum lipids by the International System of Units (SI), we converted those to conventional units with a conversion ratio of 38.6 (1 mg/dL = 0.0259 mmol/L) for TC, LDL-C, and HDL-C or 88.5 (1 mg/dL = 0.0113 mmol/L) for TG.The Newcastle-Ottawa Scale (NOS) was used to evaluate the quality of the included observational studies . A ratinI2 statistics, and a value of I2 above 50% or P < 0.1 indicated substantial heterogeneity (P < 0.05). Sensitivity analysis was performed by removing one study at a time or switching analysis model to check the robustness of the results .Detailed information on the literature search could be seen from These 15 studies were published between 1994 and 2019. Three of the studies were performed primarily in Americas , 15, fivI2= 70.1%, P = 0.000)  .P interaction > 0.05)  .P = 0.749) nor Begg's test (P = 0.876) detected publication bias. Sensitivity analysis results are in the range from 0.84(95% CI 0.70–1.00) to 0.96(95% CI 0.77–1.19) by deleting one related study each time, suggesting the stability of our meta-analysis.The funnel plot was visually asymmetry . HoweverP = 0.732), and there was no linear dose-risk relationship between them (P = 0.413) with substantial heterogeneity (P = 0.000)  .I2= 52.2%, P = 0.079)  .In subgroup analyses according to study location, statin use, and gender, similar significant inverse relationships between serum LDL-C and PD risk were observed for studies confined to a location in Americas , statin adjustment , or male . Besides, the substantial heterogeneity disappeared in the male group. We summarized the detailed results in P = 0.081) and Begg's test (P = 0.221) did not detect publication bias. Sensitivity analysis results were in the range from 0.67(95% CI 0.46–0.99) to 0.78(95% CI 0.62–0.98), suggesting that exclusion of any study did not substantially affect the pooled results. We also found a similar pooled result  using a fixed-effects model.Both Egger's test (P = 0.666). The RR of PD per 38.6 mg/dL (1mmol/L) increase in serum LDL-C concentration was 0.93 (95% CI 0.88–0.99), with little heterogeneity in the dose-response analysis (P = 0.084)  .I2= 74.5%, P = 0.003)  , suggestAs all included studies were of high quality, we conducted subgroup analyses following study location, statin use, and gender. The pooled risk estimate remained non-significant, and between-study heterogeneity decreased in either Europe or the male group .P = 0.308) and Begg's test (P = 1.000). Sensitivity analysis results range from 0.89 (95% CI 0.70–1.13) to 1.08 (95% CI 0.76–1.55) by omitting one study in order, confirming that the original results were reliable.There was no indication of publication bias assessed with Egger's test . Due to limited included studies, we could not perform subgroup analysis and evaluate publication bias.For the TG group, we obtained three cohort studies , 13, 36 Sensitivity analyses indicated little change in the pooled RR when any study was removed (data not shown).In this meta-analysis, we summarized the current evidence on the association between serum lipid parameters and the risk of PD, and thus provides a unique perspective to evaluate the effects of different serum lipid parameters on PD risk. There are some notable strengths deserving mention. Firstly, four serum lipid parameters, including TC, LDL-C, HDL-C, and TG, were included in a quantitative review to investigate their effects on PD risk for the first time. Secondly, the strict inclusion criterion of providing information on serum lipid levels before PD onset avoided the possibility of reverse causality from abnormal lipid profile induced by PD. Thirdly, both the high vs. low analysis and dose-response analysis were conducted, which ensures the reliability of the outcome.P = 0.080). However,the small number of included studies, as well as significant heterogeneity (I2= 71%) in their meta-analysis, indicated their result was unreliable and unstable. In addition, no significant association between serum TC and PD risk in the high vs. low analysis or dose-response analysis was observed, which was in line with the prior finding by Gudala et al. (In the high vs. low analysis, we found that the exposure of higher serum LDL-C levels may decrease the future risk of PD, which is consistent with the previous data on the association between serum LDL-C and PD risk , 17, 35.a et al. . For othin vitro studies and animal models of PD (Multiple lines of biological evidence responsible for the inverse linkage between serum cholesterol and PD risk have been found. Some rodent experiments supported the protective role of cholesterol precursors in the pathogenesis of PD. For example, squalene might save the striatal neurons from toxic effects caused by 6-hydroxydopamine , and lanls of PD . Besidesls of PD , olfactols of PD , depressls of PD , 47.In contrast, some hypotheses support a positive relationship between serum cholesterol levels and PD risk. As is generally recognized, abnormal intracellular aggregation of α-synuclein has an etiological role in the occurrence and development of PD . Elevatein vitro and in vivo experiments (While exploring serum lipids in relation to the risk of PD, we should draw more attention to the confounding effect of lipid-lowering drugs on this association. Statins are cholesterol-lowering drugs widely prescribed to prevent and treat cardiovascular and cerebrovascular diseases. Some studies indicated that statins might have anti-inflammation activity  and sloweriments , 58. Fureriments , 59 suggFor this paper, we should consider several limitations. Firstly, as mentioned before, altered serum lipid levels might represent a prodromal symptom of PD. Without taking into consideration the duration of the prodromal period, while still challenging to ascertain, we include the studies using data of serum lipid levels measured just before the onset of PD, which might make the results somewhat biased. Secondly, a limited number of the included studies for the relationship between serum TG and PD risk urges us to interpret its negative results with caution. Thirdly, there are some missing data in the dose-response analysis, and the approximate values inferred by existing methods may not be accurate, so the results of the dose-risk relationship might demand more sufficient data.Our meta-analysis showed that higher serum LDL-C might be a protective factor for PD and somehow reduce the future risk of PD. From an overall perspective, the protective effect of high serum LDL-C on PD risk attends to trifles and neglects essentials. After all, the increasing threat of cardiovascular and cerebrovascular events induced by hypercholesterolemia is a major public health concern. However, this study still has some implications for exploring the role of cholesterol in the pathogenesis of PD.All datasets generated for this study are included in the article/ZJ, XX, and RO designed the study and did the literature review. ZJ did the statistical analysis, edited tables and pictures, and wrote the primary manuscript. XG contributed to revising the grammar. XL helped to promote the methodology. HS and WS reviewed the final manuscript. All authors contributed to the article and approved the submitted version.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.


Obstructive sleep apnea syndrome (OSAS) is associated with a variety of systemic diseases. Among patients with chronic kidney diseases (CKD), the prevalence of OSAS is high. OSAS can induce progression of CKD. However, whether or not OSAS can cause renal damage in healthy people is not clear. Thus, the purpose of this meta-analysis was to elucidate whether or not there was an association between OSAS and early renal damage.PubMed, Embase Database, Cochrane Library, Web of Science, China National Knowledge Infrastructure, China Biology Medicine Database, Chinese Scientific Journals Database, and Wanfang Database were searched systematically. The relative risk (RR), weighted mean difference (WMD), and 95% confidence intervals (CI) were used to evaluate the relationship between OSAS and early renal damage. Funnel plot and Egger’s test were used to evaluate publication bias, and trial sequential analysis (TSA) was employed to verify the sufficiency of the research conclusions.P < 0.01) and proteinuria in patients with OSAS were significantly increased, while the levels of estimated glomerular filtration rate (eGFR)  were significantly decreased. Furthermore, patients with OSAS also had an increased risk of CKD. Subgroup analysis showed that compared with patients without OSAS, the level of serum cystatin C in patients with OSAS was significantly increased independent of hypertension and diabetes, and the eGFR was significantly decreased in patients with moderate to severe OSAS and comorbid hypertension and/or diabetes.A total of 18 studies were analyzed comprising 4,567 participants. Compared with the healthy control group, levels of cystatin C (MD = 0.530, 95% CI 0.423, 0.637, In this meta-analysis, OSAS was associated with a higher risk of early renal damage. Patients with OSAS and comorbid hypertension and/or diabetes appear to suffer from severe renal damage. Obstructive sleep apnea syndrome (OSAS) is one of the common forms of sleep disorders, characterized by recurrent episodes of apnea and hypopnea due to obstruction of the upper airway during sleep . EvidencSerum cystatin C and urinary microalbuminuria are much earlier and sensitive biomarkers for renal impairment . EmerginThis meta-analysis was conducted based on Preferred Reporting Items for Systematic Review and Meta-Analyses (PRISMA) checklist .PubMed, EMBASE Database, Cochrane Library, Web of Science, China National Knowledge Infrastructure (CNKI), China Biology Medicine Database (CBM), Chinese Scientific Journals Database (VIP), and Wanfang Database were independently searched by two researchers (from database inception to December 2019). The MeSH and keywords adjusted according to the characteristics of different databases were as follows: obstructive sleep apnea syndrome, sleep apnea, apnea-hypopnea index (AHI), sleep-associated breathing disorder, OSAS, OSA, SA, renal damage, renal dysfunction, serum cystain C, and microalbuminuria. This study includes English and Chinese. The flowchart for the process of literature selection is shown in Fig. Clinical studies eligible for this meta-analysis included cohort studies and case-control studies. Studies were considered eligible if they (1) enrolled participants with OSAS, (2) enrolled participants without CKD, (3) compared the risk of renal damage with healthy control group, and (4) provided information on any of the primary or secondary outcomes.Two researchers (HM MAO and Q LI) screened the literature independently according to the inclusion and exclusion criteria, and rescreened and extracted data for the remaining retrieval results. The literature in dispute or unable to extract data needs to be discussed with the third researcher (YL ZHAN). The content of the extracted data includes the following: the basic information, the sample size of participants, the assessment tools of OSAS, and the outcome index. The Newcastle–Ottawa Scale (NOS) provided by Cochrance was used to evaluate the quality of the included studies.2), and (3) progression of albuminuria-related indexes .The primary outcome of this meta-analysis was serum cystatin C. The secondary outcomes were (1) eGFR (as assessment by study authors), (2) incidence of CKD , so as to correct and reduce the random error.The meta-analysis was carried out with STATA 14.0, and the trial sequential analysis (TSA) was carried out with TSA0.9. If the study data was enumeration data, the relative risk (RR) and 95% confidence intervals (CI) were used as the statistical effect quantity; otherwise, the standardized mean difference (SMD) and 95%CI were used as the statistical effect quantity to evaluate the relationship between OSAS and renal damage. Chi-squared test and IA total of 3097 related literatures were retrieved. After removing the repetition, reading title and abstract, and further reading the full text, 18 cross-sectional studies , 15–30 wP < 0.01). The results of subgroup analysis showed that the serum cystatin C increased significantly in mild, moderate and severe OSAS The level of cystatin C was positively correlated with the severity of OSAS, whether they have hypertension and diabetes or not . The results of subgroup analysis showed that the decrease of eGFR is related to moderate and severe OSAS. The levels of eGFR decreased significantly in OSAS patients with hypertension and diabetes, but this phenomenon was not found in those who were not diagnosed with hypertension or diabetes.   Fig. .Fig. 4FoAdditionally, 7 studies , 22, 24 Funnel plot and Egger’s test were used to analyze the publication bias of the main indexes. Except for the subgroup of cystatin C in severe OSAS and cystatin C in OSAS without diabetes, there was no significant publication bias in other subgroup Table . In ordeWe conducted a TSA of seven studies on the relationship between cystatin C and OSAS We set type I error as 5%, and the power as 80%. Based on this analysis, the accumulated Z-curve crossed the traditional and TSA thresholds and reaches the RIS 1106) [38]. Chou . Chou  found a 8]. Chou  found thOSAS may result in the occurrence and development of CKD by aggravating traditional risk factors such as hypertension, diabetes, and obesity . On the Findings from this meta-analysis suggest that the levels of serum cystatin C and proteinuria in patients with OSAS were significantly higher, and the levels of eGFR were obviously lower than those in healthy people without OSAS. Similarly, compared with mild to moderate OSAS patients, the levels of serum cystatin C, proteinuria, and eGFR in patients with severe OSAS showed the same changes as the above results. Furthermore, our meta-analysis also found an increased risk of CKD in OSAS patients.Hypertension and diabetes are common risk factors in OSAS and CKD. In order to reduce the potential impact of hypertension and diabetes on this study, we conducted a subgroup analysis of serum cystatin C and eGFR based on whether the study included patients with hypertension and diabetes. The results showed that with or without hypertension and diabetes, the levels of cystatin C increased in OSAS patients, and there was no significant difference between the two groups. Conversely, for eGFR, OSAS with hypertension and diabetes decreased significantly.Serum cystatin C and urinary microalbumin are considered to be more sensitive markers in the diagnosis of early renal damage. Viazzi  reportedThere are some limitations in this study. First, the amount of literature included in this study is small, which may affect the accuracy of our analysis results. Second, some studies have found that obesity and male are more likely to be suffering from OSAS , while BThis meta-analysis found that OSAS was associated with a higher risk of early renal damage, especially in the patients with hypertension and/or diabetes. The result of TSA verifies the credibility and stability of the conclusions. Patients with OSAS should evaluate renal function regularly aiming to decrease the risk of developing into CKD. Notably, this meta-analysis requires more prospective cohort studies to reduce the impact of confounding factors, so as to clarify whether OSAS is an independent risk factor for renal damage.


ICAM‐1) K469E and megakaryoblastic leukaemia factor‐1 (MKL‐1) −184C/T polymorphisms with the susceptibility to coronary heart disease (CHD) in the Chinese Han population. 100 CHD patients and 91 healthy people that had no blood connection with each other were enrolled in this case‐control study. ICAM‐1 and MKL‐1 polymorphisms were genotyped by polymerase chain reaction‐restriction fragment length polymorphism (PCR‐RFLP) approach. Multiple logistic regression was used to analyse the correlation between polymorphisms of ICAM‐1 and MKL‐1 and CHD susceptibility. Differences of genotype and allele frequencies of the two SNPs between case and control groups were analysed by chi‐square test. Odds ratios (ORs) and 95% confidence intervals (CIs) were indicated relative susceptibility of CHD. The distributions of ICAM‐1 and MKL‐1 polymorphisms in each group conformed to Hardy‐Weinberg equilibrium (HWE). After adjusting for traditional risk factors, the TT genotype frequency of MKL‐1 −184C/T polymorphism was found significantly higher in case group than in control group (P < .05). Meanwhile, T allele frequency increased in case group compared with control group, and the differences had statistical significance . Logistic regression analysis in this study proved that smoking, hypertension, diabetes and triglyceride (TG) were all risk factors for CHD ICAM‐1 K469E polymorphism has no association with the onset of CHD. But MKL‐1 −184C/T polymorphism is associated with the risk of CHD and T allele might be a susceptibility factor for CHD.This study was aimed to explore the correlation of intercellular adhesion molecule‐1 ( The literature has reported that MKL‐1 might influence the formation and maintaining process of the cardiovascular system by regulating the abnormal proliferation of smooth muscle cells ICAM‐1 and MKL‐1 and CHD were inconclusive and contradictory.Coronary heart disease is a complex disease. Its specific pathogenesis is not yet fully understood. But most scholars believe that the onset of the disease is closely related to environmental and genetic factors.ICAM‐1 K469E and MKL‐1 −184C/T polymorphisms with CHD susceptibility, and explored the effect of other environmental exposures on the onset of CHD.Therefore, in the present study, we discussed the correlation of 22.1The case group of the study enrolled 100 CHD patients including 58 males and 42 females. The mean age of the patients, who were diagnosed in Cardiology department of The First Affiliated Hospital of Jinzhou Medical University, was 59.4 ± 7.33 years old. Coronary angiography examinations of every patient showed that more than one main coronary artery of the cases had 50% or higher level of stenosis. Patients suffered from cardiomyopathy, haemorrhagic diseases, renal failure and malignant tumours were eliminated. The control group recruited 91 homochronous healthy persons  from the physical examination centre of the same hospital, with a mean age of 60.3 ± 6.79 years old. The controls had no malignant tumours and immune inflammatory diseases, and they were proved without CHD history by electrocardiograph (ECG) examinations. All of the subjects were Chinese Han population, and they had no blood relationship with each other. Informed consents were obtained from each participant. And the study was approved by the Ethics Committee of The First Affiliated Hospital of Jinzhou Medical University.2.2We collected 2 mL fasting venous blood of each subject and put the blood into PCR tubes, anticoagulated by EDTA. Genome DNA was extracted from 300 μL whole blood by Biospin Whole Blood Genomic DNA Extraction Kit (Sangon Company) and then stored at −70°C.2.3ICAM‐1 K469E and MKL‐1 −184C/T polymorphisms were examined by polymerase chain reaction‐restriction fragment length polymorphism (PCR‐RFLP) approach. Primer sequences for ICAM‐1 K469E and MKL‐1 −184C/T polymorphisms were designed by Primer Premier 5.0 and synthesized by Sang Biotech  and 95% confidence intervals (95% CIs). Multiple logistic regression was adopted for testing the effects of other CHD risk factors. The differences had statistical significance when P < .05.Data analysis was performed by SPSS.18 statistical software. The genotype and allele frequencies in two groups were compared by χ33.1P > .05). But the case group had more smokers and patients with hypertension and diabetes, as well as higher levels of glucose (Glu), serum total cholesterol (TC) and triglyceride (TG) than the control group (P < .05). The differences of high‐density lipoprotein cholesterol (HDL‐C) and low‐density lipoprotein cholesterol (LDL‐C) levels between two groups were not statistically significant (P > .05).As shown in Table 3.2ICAM‐1 K469E and MKL‐1 −184C/T polymorphisms in case and control groups are shown in Table ICAM‐1 K469E polymorphism was proved to have no relation with the onset of CHD. While the distribution differences of −184C/T polymorphism TT genotype between case and control groups were of statistical significance (P < .05), logistic regression analysis offered a same result. Compared with CC genotype, TT genotype increased 4.92 times the onset risk of CHD (95% CI = 1.32‐26.6). Meanwhile, T allele was closely related to CHD as well . Thus, we confirmed that TT genotype and T allele were susceptible factors for CHD. Other related risk factors like smoking, hypertension, diabetes and TG level were consistent with the recognized CHD risk factors that had been studied before.The genotype and allele distributions of 4Coronary heart disease is a complex disease. Multiple gene mutations and risk factors lead to the occurrence and development of such disease.MKL‐1 −184C/T polymorphism with the onset of CHD in 2009, and the association was replicated in both Japanese  and Korean  populations.MKL‐1 might play important roles in the occurrence and development of atherosclerosis.MKL‐1 gene was closely related to the onset of CHD, and TT genotype increased the onset risk of such disease. T allele frequency was significantly higher in case group than that in control group, and it was suggested to be a risk factor for CHD. The distributions of TT genotype and T allele in case and control groups still had statically significant differences after logistic regression analysis, which attested the correlation with the onset of CHD. This is consistent with a previous study, and they also found that the ‐ 184C > T polymorphism of MKL1 is an important risk factor for CHD in Han nationality of Henan Province. The homozygosity of T allele is related to the risk of CHD and the severity of stenosis.With the deepening of the researches and development of biological technology, a growing body of evidence has confirmed the genetic susceptibility for CHD, and multiple genes have been found to be closely associated with the susceptibility for CHD. Hinohara et al firstly ascertained the correlation of ICAM‐1 polymorphisms and inflammatory diseases have been reported.ICAM‐1 K469E polymorphism and CHD have been reported as well both in domestic and abroad,ICAM‐1 K469E polymorphism was found to have no obvious association with CHD in the Chinese population.Many researches on the relationship between Some limitations of our research cannot be ignored. First of all, in the case‐control study, our sample is relatively small, it is necessary to expand the sample size to further verify our results. Secondly, our research objects are all Han nationality, the possibility of different genetic background and surroundings caused by ethnic and regional differences might affect experimental results. Therefore, a lager or different population should be taken into account to confirm the results.MKL‐1 −184C/T polymorphism and CHD, and T allele might be a susceptible factor for the disease. Meanwhile, the coexistence of TT genotype with smoking status, hypertension, diabetes or high TG level would significantly increase the morbidity of CHD.In summary, the present study suggested the correlation between None.Cungang Wu: Conceptualization ; Data curation ; Formal analysis ; Writing–review & editing . Chao Huang: Methodology ; Resources ; Software ; Writing–original draft .With the approval of The First Affiliated Hospital of Jinzhou Medical University Ethics Committee, written informed consent was obtaining from every subject.


Evidence from previous voxel-based morphometry (VBM) studies indicates that widespread brain regions are involved in Parkinson’s disease with mild cognitive impairment (PD-MCI). However, the spatial localization reported for gray matter (GM) abnormalities is heterogeneous. The aim of the present study was to quantitatively integrate studies on GM abnormalities observed in PD-MCI in order to determine whether a pattern exists. Eligible whole-brain VBM studies were identified by a systematic search of articles in PubMed and EMBASE databases spanning from 1995 to January 1, 2019. A meta-analysis was performed to investigate regional GM abnormalities in PD-MCI. The anisotropic effect size version of seed-based d mapping (AES-SDM) meta-analysis was conducted to explore the GMV differences of PD-MCI compared with PD patients with normal cognitive function (PD-NC). A total of 12 studies comprising 243 PD-MCI patients and 326 PD-NC were included in the meta-analysis. PD-MCI patients showed a robust GM decrease in the left insula and left superior temporal gyrus. Moreover, meta-regression analysis demonstrated that age, PD duration and stage, and Unified Parkinson’s Disease Rating Scale III and Mini-Mental State Examination scores might be partly correlated with the GM abnormalities observed in PD-MCI patients. The convergent findings of this quantitative meta-analysis revealed a characteristic neuroanatomical pattern in PD-MCI. The findings provide some evidence that MCI in PD may result in the breakdown of the insula and temporal gyrus, which may serve as specific regions of interest for further investigations. Parkinson’s disease (PD) is the most common chronic neurodegenerative disease and may lead to mild cognitive impairment (MCI) and dementia . The estNeuroimaging is useful for investigating brain structural features of PD-MCI patients. Voxel-based morphometry (VBM) is a fully automated, quantitative magnetic resonance imaging (MRI)-associated processing method for detecting subtle morphological and neuropathological changes in the whole brain and is used to quantify changes in gray matter volume (GMV) in neuropsychiatric and neurological disorders such as PD and PDD . VBM hasAnisotropic effect size-based signed differential mapping (AES-SDM) is a newly developed meta-analysis tool for neuroimaging data that has been applied to examinations of multiple disorders, including obsessive-compulsive disorder and bipolar disorder, and others –16, as Systematic and comprehensive searches of articles in PubMed and EMBASE databases published between 1995 and January 1, 2019 were performed using the following keywords: “voxel” or “VBM or morphometry or gray matter or grey matter” AND “cognitive impairment or mci or cognitive decline or cognition” AND “Parkinson’s disease or Parkinson OR PD”. A manual search of the reference lists of relevant articles was performed to identify additional potential studies.Studies were included in the meta-analysis if they met the following criteria: 1) the patient group included PD-MCI or PD-NC patients; 2) VBM studies reported a voxel-wise comparison of GM density or volume between PD-MCI and PD-NC patients; 3) whole-brain results in three-dimensional coordinates  of changes in standard stereotactic space  were reported; 4) significance thresholds corrected for multiple comparisons or uncorrected with spatial extent thresholds were used; 5) the sample size in each group was more than 5; and 6) studies were published in English as a peer-reviewed article (not as a letter or abstract). The exclusion criteria were as follows: 1) sufficient data for the meta-analysis could not be obtained even after corresponding with the authors by email; 2) findings were based solely on small volume correction; and 3) region of interest methods were used. Three researchers independently performed the study selection and data extraction and any discrepancies were resolved by consensus through discussion. The study followed the guidelines of preferred reporting items for systematic reviews and meta-analyses .http://www.sdmproject.com), a new version of SDM that has been described in detail elsewhere .Meta-regression analysis was performed to determine the correlation between clinical variables  and GMV changes that could contribute to the heterogeneity of the results. We used a voxel threshold of P<0.0005 and cluster extent=10 voxels in the meta-regression analyses. Maps showing statistical significance were visualized using MRIcron  use of a regions of interest analyses; 2) no available coordinates; and 3) published as a meta-analysis. Thus, 12 studies comprising 243 PD-MCI and 326 PD-NC patients were ultimately included in the meta-analysis –28.Voxel-based AES-SDM meta-analysis identified the temporal-insular-frontal lobe as the primary site of GM abnormalities associated with PD-MCI . RegionsSystematic whole-brain jackknife sensitivity analysis of the PD-MCI findings revealed a significant reduction in GM in the left insula in 11 of 12 datasets. GM abnormalities in the left superior temporal gyrus were highly replicable and remained significant in 10 of 12 datasets . None ofMeta-regression analyses revealed that age was negatively associated with GMV in the left rolandic operculum and that longer disease duration was correlated with greater GM atrophy in the right inferior temporal gyrus, left temporal pole, superior temporal gyrus, and left middle temporal gyrus. Additionally, higher UPDRS-III score and Hoehn and Yahr stage were correlated with greater GM atrophy in the left inferior temporal gyrus, right inferior temporal gyrus, left inferior frontal gyrus, and right middle frontal gyrus. There was no significant association between GMV reduction and education level or MMSE scores in PD-MCI patients. Detailed information is summarized in Our modified and comprehensive meta-analysis of 12 case-control VBM studies by AES-SDM showed that PD-MCI patients had less GM in the left superior temporal gyrus, left inferior frontal gyrus, and left insula. Moreover, the meta-regression analysis revealed that age, disease duration, UPDRS-III score, and Hoehn and Yahr stage were partly correlated with GM abnormalities in this group.One of the most important findings of this study was that PD-MCI patients consistently showed significant reductions in GM in the left superior temporal gyrus, left inferior frontal gyrus, and left insula compared to PD-NC patients. The results of the meta-regression analysis indicated that decreased GM in these regions in PD-MCI patients was associated with older age, longer disease duration, and higher UPDRS-III score and Hoehn and Yahr stage. Imaging studies of cognitive impairment have shown GM reduction in fronto-temporal and posterior cortical areas ,29–31. The insula and temporal gyrus constitute the ventral stream of fronto-parietal pathways that play an important role in working memory and executive function . This stInterestingly, we did not consistently observe GM abnormalities in the hippocampus, which is a hallmark of Alzheimer’s disease  even thoOur study had several limitations. First, the sample sizes of included studies were relatively small, and therefore the results must be interpreted with caution. On the other hand, a large number of studies were included and our findings can be considered robust because nearly all of the studies contributed to the results. Second, the heterogeneity of VBM study methodology including different preprocessing protocols and statistical thresholding methods may have influenced the results. Third, the results of the meta-regression analysis also found longer disease duration was correlated with greater GM atrophy in the left superior temporal gyrus. This seems to suggest the degenerative course of the disease in PD also may have occurred along grey matter loss in the left superior temporal gyrus, which may have led to less accurate results. However, a recent meta-analysis found that GM reductions were detected in the right temporal lobe, right rolandic operculum, right amygdala, right angular gyrus, right middle occipital gyrus, right fusiform gyrus, left superior frontal gyrus, bilateral anterior cingulate/paracingulate gyri, bilateral insula, and bilateral striatum in the PD patients compared with the healthy controls. Regions with significantly less GM were the left insula, left superior temporal gyrus, and left inferior frontal gyrus. No significant regional GM atrophy in left superior temporal gyrus was detected in the PD patients .In conclusion, the present voxel-wise meta-analysis revealed that GM was reduced in the left insula and left superior temporal gyrus of PD-MCI patients compared with PD-NC patients. Furthermore, we found that age, disease duration, UPDRS-III scores, and Hoehn and Yahr stage were partly correlated with GM abnormalities in PD-MCI patients. These findings may provide a basis for investigations on the mechanisms underlying changes in the visual cortex that are associated with cognitive decline in PD. Future longitudinal and multicenter MRI studies should investigate whether this brain morphometric pattern can serve as a useful target and a prognostic marker for PD-MCI diagnosing and treatment.


PRSS1-PRSS2 haplotype and alcoholic chronic pancreatitis (ACP), which was revealed by the first genome-wide association study of chronic pancreatitis (CP), has been consistently replicated. However, the association with non-ACP (NACP) has been controversial. Herein, we sought to clarify this basic issue by means of an allele-based meta-analysis of currently available studies. We then used studies informative for genotype distribution to explore the biological mechanisms underlying the association data and to test for gene-environment interaction between the risk haplotype and alcohol consumption by means of a re-analysis. A literature search was conducted to identify eligible studies. A meta-analysis was performed using the Review Manager software. The association between the risk genotypes and NACP or ACP was tested for the best-fitting genetic model. Gene-environment interaction was estimated by both case-only and multinomial approaches. Five and eight studies were employed for the meta-analysis of ACP and NACP findings, respectively. The risk allele was significantly associated with both ACP (pooled odds ratio (OR) 1.67, 95% confidence interval (CI) 1.56–1.78; p < 0.00001) and NACP . Consistent with a dosage effect of the risk allele on PRSS1/PRSS2 mRNA expression in human pancreatic tissue, both ACP and NACP association data were best explained by an additive genetic model. Finally, the risk haplotype was found to interact synergistically with alcohol consumption.The association between a common  PRSS1 gene  can cause autosomal dominant hereditary pancreatitis [Chronic pancreatitis (CP) is a chronic inflammatory process of the pancreas that leads to irreversible morphological changes and progressive impairment of both exocrine and endocrine functions ,2. CP careatitis ,5. In threatitis .PRSS1-PRSS2 haplotype was reported to be associated with both alcoholic CP (ACP) and non-ACP (NACP) [PRSS2 encodes anionic trypsinogen (MIM# 601564), the second major trypsinogen isoform after cationic trypsinogen. This association has a strong biological basis: the risk (C) allele of the lead single nucleotide polymorphism (SNP), rs10273639C/T, appeared to be associated with increased PRSS1 mRNA expression in pancreatic tissue in a dosage-dependent manner [PRSS1 gene; the C allele of rs4726576C/A increased reporter gene expression as determined by an in vitro promoter assay [Eight years ago, the first genome-wide association study of CP was published; a common P (NACP) . PRSS2 egen MIM# 01564, ther assay .PRSS1-PRSS2 haplotype with ACP has been consistently replicated. However, the association with NACP was equivocal, raising the possibility that the common PRSS1-PRSS2 haplotype modifies CP risk only for ACP but not in NACP [PRSS1-PRSS2 haplotype and ACP and NACP. We then used informative studies in terms of genotype distribution to perform a new re-analysis with respect to (i) the fit of different genetic models to the association data and (ii) the gene-environment interaction between the common PRSS1-PRSS2 haplotype and alcohol consumption status in CP.Another interesting finding from the Whitcomb study was that rs10273639C/T was associated with a stronger genetic effect in ACP than in NACP, suggesting that the SNP-conferred risk was amplified by alcohol consumption. Moreover, in a case-only analysis, the SNP seemed to interact with ACP [PRSS1 and Pancreatitis) via PubMed (as of 13 July 2020), which was complemented by the perusal of references citing the Whitcomb and/or Derikx studies [PRSS1-PRSS2 haplotype was analysed in both ACP and/or NACP patients and controls; and (iii) the tagging SNP of interest was in Hardy–Weinberg equilibrium in the controls.The literature search was performed by keyword search  that tagged the common 2 tests in R [p-value was ≤0.05. The sample size for case-control association studies was estimated using the Online Sample Size Estimator [The assessment of the significance of the differences between the frequencies of the risk-tagging SNPs in patients and controls in the context of each study was performed by means of χsts in R . A diffestimator .p-value for the test of heterogeneity was <0.05 or I2 was 50% or more [Meta-analysis, heterogeneity and sensitivity analyses and forest plots were all performed using the Review Manager 5.3 software . Funnel  or more ,15. The  or more . This me or more , which dThis analysis was performed using the Genotype-Tissue Expression (GTEx) dataset .Logistic regression models were fitted to examine the association between the risk genotypes and NACP or ACP in the context of single population data (refer to Results). Four models were tested: the genotypes were coded as either recessive, dominant, additive, or as each having an independent effect  . We evalThe studies included in this analysis were those for which we could ascertain individual-level genotype data for ACP, NACP and healthy control individuals. We firstly performed case-only analyses before combining results through meta-analysis in order to estimate a pooled OR for the interaction term . This analysis was performed using the metagen function from the meta R package .In a second step, we applied the multinomial approach to test for gene-environment interactions as proposed by Kazma and colleagues . ACP andPRSS1 and Pancreatitis) via PubMed resulted in 173 entries starting from the pioneering Whitcomb study [PRSS1-PRSS2 haplotype in both pancreatitis patients and controls. Two studies that analysed patients with a known non-alcohol-related external factor [The keyword search . The pooled OR (under a fixed effect model) was 1.67   between the American (primarily of European ancestry), European, Japanese, Indian and Hungarian control populations studied, all five ACP studies found a statistically significant association between the common 0.00001) .2 = 15.81, p = 0.03; I2 = 56%). Sensitivity analysis identified the large European study [PRSS1-PRSS2 haplotype showed significant association with NACP: the pooled OR was 1.28   .p = 0.17) in the corresponding study [The three studies that failed to show a significant association with NACP nevertheless, all yielded ORs > 1 . Notablyng study  might hap-value of 0.085  allele of rs10273639C/T is associated with increased trypsinogen gene expression in the pancreas in the GTEx dataset. No significant expression quantitative trait loci (eQTLs) for t manner .PRSS1 and PRSS2 mRNA expression implies that an additive genetic model would best explain the association data. To test this postulate, we firstly examined the three possible genetic models , in the context of three NACP studies informative for genotype distribution, each from a single population   .We also examined the three possible genetic models in the context of ACP studies. Of the five included ACP studies , four usPRSS1/PRSS2 haplotype. Following the evaluation of different genetic models in NACP and ACP patients, we selected the best-performing additive genetic model for this re-analysis. Subsequently, we performed a meta-analysis of these new results in order to calculate a pooled estimate for the interaction term and to assess the heterogeneity of the studies. Application of the test for heterogeneity confirmed that the five studies were homogeneous . The pooled OR (under a fixed effect model) of the interaction  was 1.41 . That the OR is significantly greater than one indicates that the interaction effects between alcohol exposure and the haplotype are more than multiplicative (synergistic effect).Using data from ACP and NACP patients from Germany, France, Netherlands, Japan and Hungary , we firsPRSS1-PRSS2 haplotype in both ACP and NACP cases (PRSS1/PRSS2 haplotype was 1.16 (95% CI 1.07–1.26) in the NACP patients and increased to 1.69 (95% CI 1.54–1.84) in the ACP patients. The difference between these two ORs was statistically significant, as evidenced by the non-overlapping confidence intervals, again signalling the presence of an interaction between the genetic-effect of the haplotype and alcohol consumption. This model allows for a succinct demonstration that alcohol consumption significantly increases the risk associated with the presence of copies of the haplotype on a multiplicative scale across the five populations.To fully characterise the importance of the interaction, we fitted a multinomial logistic regression model to estimate, within a single framework, the risks associated with the CP cases . For thiPRSS1-PRSS2 haplotype and ACP and NACP, demonstrating a significant association in both contexts. Possible causes underlying the previously contentious NACP findings included small sample size, ethnic and population differences in genetic predisposition to CP, and differences in patient selection. For example, using only German samples, Derikx and colleagues observed a trend towards higher genetic effect sizes in patient groups characterised by a later age of disease onset [PRSS1-PRSS2 rs10273639 was included in a recent meta-analysis, in which ACP and NACP were analysed together [We first performed an allele-based meta-analysis of the currently available studies of the association between the common se onset , a possitogether .PRSS1-PRSS2 haplotype with NACP, we further explored the underlying biological basis of this association from two complementary standpoints. As mentioned earlier, the risk (C) allele of rs10273639C/T has been found to be positively associated with PRSS1 mRNA expression in pancreatic tissue. More specifically, PRSS1 expression levels were highest in rs10273639(C) homozygotes, intermediate in heterozygotes and lowest in rs10273639(T) homozygotes [PRSS1 mRNA expression analysis was limited. We, therefore, evaluated the effect of rs10273639 genotypes on pancreatic trypsinogen gene expression using the GTEx dataset and found the level of PRSS2 mRNA expression to correlate directly with the dosage of the risk (C) allele of rs10273639. A role for PRSS2 in CP is supported by four lines of evidence. Firstly, a loss-of-function PRSS2 variant is known to be protective against CP [PRSS1 and PRSS2 [CTRB1-CTRB2 inversion increases CP risk by influencing protective PRSS2 degradation [PRSS2 exacerbates pancreatitis in mice [PRSS1/PRSS2 expression in the pancreas implies that the additive gene model represents the best fit to explain the disease association data. Thus, the second approach we adopted was to test this hypothesis by a series of statistical analyses. We demonstrated that the additive genetic model provided the best fit in both the ACP and NACP contexts and the genetic effect of the rs10273639 CC genotype was approximately double that of the CT genotype. Taken together, these findings provide new insights into the pivotal role of increased trypsin in the aetiology of CP as well as the genotype-phenotype relationship in ACP/NACP.Having confirmed an association of the common nd PRSS2 ,40,41. Tradation . Finally in mice . A dosagPRSS1-PRSS2 risk allele associated with ACP than with NACP suggests a gene-environment interaction between the common haplotype and alcohol consumption in CP. By comparing allele distributions at rs10273639 between 447 ACP patients and 1129 NACP patients, Whitcomb and colleagues [PRSS1-PRSS2 haplotype and alcohol consumption although they did not attempt to quantify it. Herein, by fitting a similar but more sophisticated model, and by combining inference from data of several studies, we confirm that there is indeed a synergistic interaction between the PRSS1-PRSS2 haplotype and alcohol consumption status, i.e., the risk associated with the presence of the PRSS1-PRSS2 haplotype and exposure to alcohol was significantly greater than the product of the risks associated with each factor singly. Our results are in accordance with those of Whitcomb and colleagues but here, by first determining the correct genetic model (additive), and second by applying multinomial logistic regression, a complete characterisation of the aetiology of CP regarding the PRSS1-PRSS2 haplotype and alcohol consumption was obtained. These findings have important implications for risk assessment and disease prevention in the age of precision medicine and should also stimulate new studies designed to establish a general framework for understanding gene-alcohol interactions in CP.Alcohol is an established environmental risk factor for CP . HoweverThis study has several limitations. For example, we could not exclude publication bias in terms of some of the included studies. We did not perform funnel plots to assess this possibility following advice that such a test should be used only when there are at least ten eligible studies; otherwise, the power of the test would have been too low to distinguish a chance effect from genuine asymmetry ,14. DespPRSS1-PRSS2 haplotype with both ACP and NACP. In addition, we have provided new information that sheds light on the pivotal role of uncontrolled trypsin in CP. Finally, we have refined the gene-environment interaction between the common PRSS1-PRSS2 haplotype and alcohol consumption status in the aetiology of CP. These findings help to improve our understanding of the complex aetiology of CP and strengthen the notion that common risk factors [PRSS1-PRSS2 haplotype should consider greatly reducing their alcohol consumption or abstaining altogether.In conclusion, the results of our allele-based meta-analysis demonstrated a significant association of the common  factors ,30,44 sh