Introduction: Association between Cyclin D1 (CCND1) single nucleotide polymorphism (SNP) rs9344 and cancer risk is paradoxical. Thus, we performed a meta-analysis to explore the association between CCND1 variant and overall cancer risk in Indian population. Methods: Data from 12 published studies including 3739 subjects were collected using Pubmed and Embase. RevMan (Review Manager) 5.3 was used to perform the meta-analysis. OR with 95%CI were calculated to establish the association. Results: Overall, the cumulative findings demonstrated that CCND1 polymorphism (rs9344) was not significantly associated with cancer risk in all the genetic models studied (dominant model: GG vs GA+AA: OR (95%CI) = 0.81 (0.60–1.09), P=0.17; recessive model: GG+GA vs AA: OR (95%CI) = 1.23 (0.96–1.59), P=0.11; co-dominant model: P=0.12;GG vs AA: OR (95%CI) = 1.35 (0.93–1.97),  co-dominant model: P=0.34(GG vs GA: OR (95%CI) = 1.16 (0.85–1.59), ; allelic model: A vs G: OR (95%CI) = 1.20 (1.14–2.85), P=0.23; allelic model: G vs A: OR (95%CI) = 0.83 (0.62–1.12), P=0.23). Subgroup analysis according to cancer types presented significant association of CCND1 polymorphism and increased breast cancer risk in dominant model  and allelic model . An increased esophageal cancer risk in recessive model  and co-dominant model  was detected. A higher risk for colorectal cancer was detected under both the co-dominant models . However, in case of cervical cancer risk a non-significant association was reported under the recessive model  with reference to CCND1 polymorphism (rs9344). The trial sequential analysis (TSA) showed that the cumulative Z-curve neither crossed the trial sequential monitoring boundary nor reached the required information size (RIS). Thus, present meta-analysis remained inconclusive due to insufficient evidence. Conclusion:CCND1 polymorphism rs9344 may not have a role in overall cancer susceptibility in Indian population. However, this polymorphism acts as a crucial risk factor for breast, esophageal, and colorectal cancer but not for cervical cancer. Future studies with larger sample size are required to draw a reliable conclusion. Cancer is a major global health problem and it is worse in case of low- and middle-income developing countries. According to India’s National Cancer Registry Program (NCRP), 1.45 million cases would occur in 2016 with 0.74 million deaths in India. This is expected to rise to 1.73 million cases and 0.88 million deaths in 2020  was first described by Betticher et al. (1995) [CCND1 and responsible for alternate splicing of transcripts with different half-lives [CCND1 SNP (rs9344) and cancer susceptibility. Occurrence of this nucleotide variation has been found to be coupled with the risk of various cancers including cervical, breast, oral, esophageal, lung, urinary bladder, prostate, and colorectal [CCND1 polymorphism and risk for different types of cancer [CCND1 SNP and overall cancer risk. Hence, we aimed to investigate the role of CCND1 polymorphism G870A (rs9344) in overall cancer susceptibility amongst Indian population by conducting this meta-analysis. The present data could be helpful in enriching the existing knowledge with respect to involvement of CCND1 polymorphism and cancer susceptibility in Indian population.. (1995) . This SNlf-lives . Since tlorectal . The outf cancer . To the Pubmed and Embase databases were searched with the keywords ‘CCND1’, ‘Cyclin D1’, ‘SNP’, ‘cancer’, ‘India’, and ‘polymorphism’ for literature published till September 2016. All studies included in the present meta-analysis met the following inclusion and exclusion criteria.CCND1 SNP G870A (rs9344) and cancer susceptibility, (ii) studies included Indian population, (iii) genotypic and allelic details are provided for both the cases and control groups, (iv) full text available, and (v) articles published in English language.(i) Prospective or case–control studies involving association analysis between (i) Studies published on populations other than Indian, (ii) articles published in languages except English, and (iii) articles not providing genotypic and allelic details.Data from all eligible studies were retrieved independently by two investigators (N.T. and S.K.). The retrieved data incorporated the following details: (i) PubMed IDentifier (PMID), (ii) name of the first author, (iii) year of publication, (iv) country, (v) sources of controls, (vi) methods for genotyping, and (vii) frequency of genotypic and allelic data.Quality of the included studies was assessed by assigning the quality scores as previously mentioned by He et al. (2014) . The scoReview Manager) is an easy tool to perform the meta-analyses and generate the graphs  in publication standard. Meta-analysis of CCND1 gene G870A polymorphism (rs9344) was performed by RevMan 5.3 [. For statistical models, both fixed model and random model were included in the RevMan. For random models, DerSimonian and Laird random-effects models were used. Odds ratios (ORs) with 95% confidence intervals (95%CIs) were used to assess the strength of association between the CCND1-G870A polymorphisms and cancer risk. The pooled OR was evaluated by the Z-test and a P-value <0.05 suggests a significant association.RevMan , recessive (AA vs GG+GA), co-dominant (GA vs GG and AA vs GG), and allelic (A vs G and G vs A). The P-value <0.05 was considered statistically significant. Subgroup analysis was done after stratification of data according to various cancer types.The association between 2I statistic. The OR of different types of genetic models was evaluated by employing the fixed-effect model (when 2I < 50%) or random-effect model (when 2I > 50%). Egger’s bias test and Begg’s funnel plot was used to assess the publication bias [http://www.ctu.dk/tsa/) [Heterogeneity was calculated by chi-square test and the extent of heterogeneity was measured by the value of ion bias ,41. It iion bias . Hence, ion bias . It was dk/tsa/)  to calcudk/tsa/) . If the Pubmed and Embase database, a total of 12 studies were searched independently by two investigators (S.K. and N.T.) according to the methodology depicted in flow diagram  and cancer risk in 12 case–control studies are shown in P=0.17, 2I= 72%); recessive model ; co-dominant model ; co-dominant model ; allele model , and allele model   is not associated with the risk for overall cancers in Indian population.A total of 12 studies were included in the analysis to evaluate the association between 2 = 82%) . If the CCND1 polymorphism and increased breast cancer risk in dominant model , allelic model . A statistically significant association with esophageal cancer risk was observed in recessive  and co-dominant model . An increased risk for colorectal cancer was detected under both the co-dominant models . Contrary to this, none of the genetic model reported a statistically significant association with cervical cancer risk. Although a non-significant association was observed in recessive model  and co-dominant model  with reference to CCND1 polymorphism (rs9344)  .2I was calculated for different genetic models and presented in 2I = 72%, P for heterogeneity <0.0001; recessive model: GG+GA vs AA: 2I = 64%, P for heterogeneity = 0.001; co-dominant model: GG vs AA: 2I = 69%, P for heterogeneity = 0.0002; co-dominant model: GG vs GA: 2I = 72%, P for heterogeneity = 0.0001; allelic model: A vs G: 2I = 82%, P for heterogeneity <0.00001 and allelic model: G vs A: 2I = 82%, P for heterogeneity <0.00001 were noted, respectively  and overall cancer risk showed that only conventional boundary was crossed by Z-curve, however, it neither crossed the TSA boundary nor the futility area. And the total sample size (3739) did not reach the RIS (11375) . This reCCND1 is key driver of normal cell cycle regulation and genetic variation in this gene has been reported in many types of cancers. A SNP G870A (rs9344) located on exon-4–intron boundary of CCND1 has been studied extensively in several cancer types. Several reports from different parts of the world have been published with reference to CCND1 polymorphism and risk of various types of cancers including cervical, prostate, colorectal, urinary bladder, squamous cell carcinoma of the head and neck etc. [CCND1 G870A (rs9344) polymorphism and overall cancer risk in Indian population. Although, various meta-analyses on individual cancer susceptibility have been published globally [CCND1 polymorphism in different types of cancers and populations. However, a comprehensive data are lacking from India with overall cancer risk. Hence, we have designed the present study focussed on Indian population.eck etc. . Investieck etc. . Howeverglobally . Pabalanglobally , performCCND1 G870A polymorphism (rs9344) and overall cancer risk in all the genetic models. These findings are consistent with the result of another study by Luo et al. (2016) [CCND1 polymorphism (G870A) with the risk of hepatocellular carcinoma. In the similar lines, study by Zheng et al. (2015) [CCND1 polymorphism may not be associated with the risk of prostate cancer. Similarly, Wang et al. (2018) [CCND1-A870G polymorphism in the human population. Another study by Qin et al. (2014) [CCND1 polymorphism may increase the risk of non-Hodgkin lymphoma but it was not true in case of leukemia. On the identical lines, Lin et al. (2014) [CCND1 polymorphism (G870A) and head and neck cancer, however; they found that smokers carrying ‘A’ allele or ‘AA’ genotype for rs9344 SNP located on CCND1 may be at higher risk to head and neck cancer development.The present meta-analysis, contained a total of 12 studies comprising 1791 cancer cases and 1948 controls  showed t. (2016) , which r. (2015)  suggeste. (2018)  also fou. (2018) , showed . (2014)  also ind. (2014)  too obseP=0.38). The possible explanation for this observation may be the small sample size of contributing studies. No promising association of this SNP has been established with the development of cervical cancer in Caucasian population by Yang et al. (2015) [CCND1 SNP (rs9344) and overall risk for cervical cancer in the Asian population but on stratification analysis by race, individuals carrying the AA or AA/AG genotypes showed a significant higher risk in comparison with GG carriers [CCND1 G870A polymorphism and cervical cancer risk amongst different ethnic groups including Asian, Caucasian, and mixed in a cumulative meta-analysis.Our subgroup analysis showed an increased risk (1.52-fold) for cervical cancer development but this association could not attain the limits of statistical significance (. (2015) . In anotcarriers . In paracarriers , also diCCND1 polymorphism and increased risk for breast and esophageal cancer has been established. Similar to our results, Sergentanis and Economopoulos (2011) [CCND1 G870A polymorphism is associated with higher risk for breast cancer. These findings are further strengthened by another meta-analyses conducted by Lu et al. (2009) [CCND1 G870A polymorphism with breast cancer susceptibility. Similarly, Soleimani et al. (2016) [CCND1 G870A polymorphism and breast cancer risk but in Caucasians. A meta-analysis conducted Wen et al. (2014) [CCND1 G870A polymorphism is a potential risk factor in the development of esophageal cancer. Other related meta-analysis by Cai et al. (2013) [CCND1 G870A polymorphism and esophageal cancer risk. Likewise, Tang et al. (2015) [CCND1 SNP rs9344 is not having role in esophageal squamous cell carcinoma.The present study suggests that there is a significant correlation between this polymorphism and increased risk of colorectal cancer amongst Indian population. Recently, Xu et al. (2016) [CCND1 G870 AA genotype may increase the colorectal cancer risk compared with the GG+AG genotype  in an Indian population. Similarly, Zhang et al. (2016) [CCND1 polymorphism is a risk factor for gastric cancer in Caucasians. According to the literature search, Dai et al. (2016) [CCND1 polymorphism (rs678653) located on the 3′-UTR and susceptibility to cancer, but they have not studied the polymorphism under investigation G870A (rs9344).Additionally, a significant association between s (2011)  found th. (2009)  and Cui . (2009)  that sho. (2016)  showed a. (2014)  supporte. (2013)  is not i. (2015) , also ob. (2016)  suggeste. (2016) , suggest. (2016)  also triThe present study had some limitations, first, all of the included studies were hospital based which may not represent the true population. Second, environmental factors like smoking, use of alcohol, and infections with viruses were not included in the present meta-analysis. Finally, the sample size was reasonably small, which may be the reason for controversial results.CCND1 SNP (rs9344) may not serve as a risk factor for overall cancer susceptibility in Indian population. However, a significant association between CCND1 SNP and increased risk for breast, esophageal, and colorectal cancer was found on subgroup analysis. Moreover, a non-significant increased risk for cervical cancer in relation to CCND1 polymorphism was observed in Indian population. Thus, CCND1 G870A (rs9344) polymorphism has a potential to be served as a prognostic biomarker for breast, esophageal, and colorectal cancer in Indian population. Still, larger and well-designed studies including other risk factors are warranted in future to validate the findings from present analysis.In conclusion, present meta-analysis showed that


In 2005, the FDA cautioned that exposure to paroxetine, a selective serotonin reuptake inhibitor (SSRI), during the first trimester of pregnancy may increase the risk of cardiac malformations. Since then, the association between maternal use of SSRIs during pregnancy and congenital malformations in infants has been the subject of much discussion and controversy. The aim of this study is to systematically review the associations between SSRIs use during early pregnancy and the risk of congenital malformations, with particular attention to the potential confounding by indication.The study protocol was registered with PROSPERO (CRD42018088358). Cohort studies on congenital malformations in infants born to mothers with first-trimester exposure to SSRIs were identified via PubMed, Embase, Web of Science, and the Cochrane Library databases through 17 January 2018. Random-effects models were used to calculate summary relative risks (RRs).Twenty-nine cohort studies including 9,085,954 births were identified. Overall, use of SSRIs was associated with an increased risk of overall major congenital anomalies  and congenital heart defects . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis . Similar significant associations were observed using maternal citalopram exposure , fluoxetine , and paroxetine  and analyses restricted to using women with a psychiatric diagnosis were not statistically significant. Sertraline was associated with septal defects , atrial septal defects , and respiratory system defects .The evidence suggests a generally small risk of congenital malformations and argues against a substantial teratogenic effect of SSRIs. Caution is advisable in making decisions about whether to continue or stop treatment with SSRIs during pregnancy.The online version of this article (10.1186/s12916-018-1193-5) contains supplementary material, which is available to authorized users. Selective serotonin reuptake inhibitors (SSRIs) have become the first-line pharmaceuticals for the treatment of depression, anxiety, and other psychiatric disorders since they were introduced into the market . About 6The number of published meta-analyses regarding the associations between maternal use of SSRIs and congenital malformations has more than tripled during the last 5 years . However, some of these studies produced conflicting results due to varying study designs and exposure times. Most inconsistently reported were congenital heart defects (CHD)  group . Before PubMed, Embase, Web of Science, and The Cochrane Library were searched from database inception to 17 January 2018. The search strategy combined medical subject heading (MeSH) and Embase subject heading (EMTREE) terms with other unindexed or free-text terms. Details of the full search strategy are provided in Additional file Cohort studies or randomized controlled trials that reported original data were eligible for inclusion if they reported any congenital malformations in infants born to mothers with any exposure to SSRIs or individual SSRIs  during the first trimester, had a comparison group that included pregnant women who were not exposed to any antidepressants and/or teratogens , and, if a risk estimate was not reported, provided necessary distribution of exposure, non-exposure, cases, and non-cases, from which a risk estimate could be calculated.The titles and abstracts of retrieved articles were evaluated by two independent reviewers (S-YG and CS). The full texts of potentially eligible studies that seemed to meet the inclusion criteria were then obtained and independently reviewed by the two reviewers. Any disagreements were identified and resolved by discussion or by consultation with a third reviewer (Q-JW). If data were duplicated in more than one study, we included the study with the largest number of cases.A standardized, pre-designed spreadsheet was used for data extraction from the included studies. The study quality and synthesis of evidence were assessed. The following data were extracted into the spreadsheet: first author, publication year, geographic location, study period, data source, sample size (cases and cohorts), types of birth, definition of outcome, outcome with their risk estimates and 95% confidence intervals (CIs), and adjusted confounders. Congenital malformations were identified and defined according to the European Surveillance of Congenital Anomalies (EUROCAT) Guide 1.3 and ICD-10 and ICD-9 codes  . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis , atrial septal defects (ASD) [I2 = 72.0%, P = 0.002), and right ventricular outflow tract defects (RVOTD) [I2 = 33.0%, P = 0.21) .Maternal use of SSRIs during the first trimester was associated with an increased risk in septal defects , 43, 72 ts (ASD) , 61, 72  (RVOTD) , 55, 72 I2 = 0, P = 0.43), cystic kidney disease [I2 = 0, P = 0.81), clubfoot [I2 = 0, P = 0.65), abdominal wall defects [I2 = 0, P = 0.86), omphalocele [I2 = 0, P = 0.73), and gastroschisis [I2 = 0, P = 0.56) , with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Eight studies , 61, 72 I2 = 52.5%, P = 0.02), with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Eleven studies , 61, 72 I2 = 0, P = 0.55), RVOTD [I2 = 0, P = 0.54), eye defects [I2 = 0, P = 0.55), urinary system defects [I2 = 0, P = 0.72), and hypospadias [I2 = 0, P = 0.43) , RVOTD , 39 , with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Eleven studies , 61, 72 I2 = 29.3%, P = 0.14), with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Fourteen studies , 61, 72  = 0.18) , 72 , RVOTD [I2 = 18.0%, P = 0.30), neural tube defects [I2 = 0, P = 0.76), and ear, face, and neck defects [I2 = 0, P = 0.41) , RVOTD , 39, 55  defects , 37, 39  defects , 40 , with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Eleven studies , 61, 72 I2 = 0, P = 0.71), with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Sixteen studies , 66, 72  = 0.03) , 72 , eye defects [I2 = 0, P = 0.53), and cleft palate [I2 = 0, P = 0.83) , with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Nine studies , 61, 72 I2 = 63.9%, P = 0.001), with no evidence of publication bias . No significantly increased risk was observed when restricted to women with a psychiatric diagnosis  [Thirteen studies , 61, 72  = 0.80) , 72 , ASD [I2 = 0, P = 0.54), respiratory system defects [I2 = 0, P = 0.45), limb defects [I2 = 0, P = 0.54), and clubfoot [I2 = 0, P = 0.77) , ASD , 39, 40  defects   after excluding the study by Huybrechts et al. [The results of subgroup and meta-regression analyses are presented in Additional file s et al. .This comprehensive systemic review and meta-analysis of cohort studies including more than nine million births found generally small increased risks in 18 of 29 congenital malformations categories in infants born to mothers with exposure to SSRIs and individual SSRIs during early pregnancy, especially for MCAs and CHD. We found the RRs for the association between use of SSRIs and outcomes were lower in the restricted cohorts. Though the 95% CIs of the comparisons made between studies in general population and studies in mothers with a psychiatric diagnosis contained an overlap, we still cannot exclude the possibility of confounding by indication.We identified a small but significant association between maternal use of SSRIs during the first trimester and MCAs in infants. This observed increase was consistent with the results of previous meta-analyses , 19 HoweThere was an association between first-trimester exposure to SSRIs and the risk of CHD. Some biological evidence possibly supports the observed increase , 73, butThe observed increase in MCAs and CHD with maternal use of citalopram during the first trimester was inconsistent with all previous meta-analyses , 14, 15.The main results of this meta-analysis of fluoxetine use during early pregnancy were consistent with our previous study  and a reOur data show a significant association between the use of paroxetine and the risks of MCAs and CHD. The results were consistent with a recent meta-analysis by Berard and colleagues , but werAlthough the findings of this meta-analysis were consistent with our previous study , our datThe main results of this meta-analysis were consistent with previous meta-analyses , 29 rega, a growing body of evidence has suggested that the neurotransmitter serotonin (5-HT) plays a crucial role as a signaling molecule in cardiogenesis [Potential biological mechanisms of SSRI use and the increased risk of congenital malformations is based on studies of drug metabolite levels in cord blood in human . In vitrogenesis , 75. Conogenesis , 76.Our study has several strengths. First, this meta-analysis of current evidence from cohort studies includes the largest sample size (more than nine million births) analyzed to date and combines the results with the most comprehensive data related to associations between the use of SSRIs and all individual SSRIs during early pregnancy and the risks of 29 categories of congenital malformations. Second, this meta-analysis pays particular attention to the potential confounding by indication. Third, for ethical reasons, there are no randomized controlled trials; therefore, the quality of evidence from cohort studies could provide clinicians and pregnant women with a reference in clinical practice.There are limitations in our meta-analysis related to evidence synthesis and quality. First, the definition of outcomes varied among studies, particularly the definition of CHD, which could contribute to the high heterogeneity in our study. CHD has a specific definition and coding in the EUROCAT guide, but not in the ICD codes; however, most of the individual studies defining outcomes were based on ICD codes such as ICD-10 and ICD-9  on system-specific malformations, specifically in women with underlying psychiatric diagnosis. The accumulated evidence suggests a generally small risk of congenital malformations and argues against a substantial teratogenic effect of SSRIs. Caution is advisable in making decisions about whether to continue or stopping treatment with SSRIs during pregnancy. Stopping treatment in mothers with major depression could be more harmful for the infant than continuing use of SSRIs. This information could be helpful for pregnant women and their healthcare providers to make more informed decisions about treatment.Additional file 1:Table S1. Characteristics of prior meta-analyses of selective serotonin reuptake inhibitors (SSRIs) use in pregnancy and congenital malformations. (DOC 816 kb)Additional file 2:Appendix 1. Search strategy. (DOCX 26 kb)Additional file 3:Table S1. EUBOCAT Guide 1.3, ICD-10, and ICD-9 codes used to identify and define congenital malformations. Table S2. Characteristics of cohort studies of selective serotonin reuptake inhibitors (SSRIs) use in first-trimester and congenital malformations. Table S3. Risk of bias of included reports from cohort studies as assessed with the Newcastle-Ottawa scale. Table S4. Exposure to selective serotonin reuptake inhibitors (SSRIs) during the first trimester of pregnancy and risk of congenital malformations in infants: results of meta-analyses. Table S5. Exposure to citalopram during the first trimester of pregnancy and risk of congenital malformations in infants: results of meta-analyses. Table S6. Exposure to fluoxetine during the first trimester of pregnancy and risk of congenital malformations in infants: results of meta-analyses. Table S7. Exposure to paroxetine during the first trimester of pregnancy and risk of congenital malformations in infants: results of meta-analyses. Table S8. Exposure to sertraline during the first trimester of pregnancy and risk of congenital malformations in infants: results of meta-analyses. Table S9. Exposure to escitalopram during the first trimester of pregnancy and risk of congenital malformations in infants: results of meta-analyses. Table S10. Exposure to fluvoxamine during the first trimester of pregnancy and risk of congenital malformations in infants: results of meta-analyses. Table S11. Subgroup analysis of selective serotonin reuptake inhibitors (SSRIs) and risk of congenital malformations in infants: results of meta-analyses. Table S12. Subgroup analysis of citalopram and risk of congenital malformations in infants: results of meta-analyses. Table S13. Subgroup analysis of fluoxetine and risk of congenital malformations in infants: results of meta-analyses. Table S14. Subgroup analysis of paroxetine and risk of congenital malformations in infants: results of meta-analyses. Table S15. Subgroup analysis of sertraline and risk of congenital malformations in infants: results of meta-analyses. (DOC 1162 kb)Additional file 4:Figure S1. Risk of major congenital anomalies in infants, according to maternal exposure to citalopram. (TIF 1083 kb)Additional file 5:Figure S2. Risk of congenital heart defects in infants, according to maternal exposure to citalopram. (TIF 1123 kb)Additional file 6:Figure S3. Risk of major congenital anomalies in infants, according to maternal exposure to fluoxetine. (TIF 1141 kb)Additional file 7:Figure S4. Risk of congenital heart defects in infants, according to maternal exposure to fluoxetine. (TIF 1177 kb)Additional file 8:Figure S5. Risk of major congenital anomalies in infants, according to maternal exposure to paroxetine. (TIF 1135 kb)Additional file 9:Figure S6. Risk of congenital heart defects in infants, according to maternal exposure to paroxetine. (TIF 1244 kb)Additional file 10:Figure S7. Risk of major congenital anomalies in infants, according to maternal exposure to sertraline. (TIF 1112 kb)Additional file 11:Figure S8. Risk of congenital heart defects in infants, according to maternal exposure to sertraline. (TIF 1185 kb)


Background: Recently, many case-control studies have reported the association between 5-HTTLPR polymorphism and autism risk. However, the results are inconclusive and conflicting. To investigate the genetic association of 5-HTTLPR polymorphism and autism risk, we conducted a comprehensive meta-analysis based on previous case-control studies.Methods: Literature search was performed through PubMed, Embase, Web of Knowledge and CNKI databases until June 27, 2018. The strength of the association was assessed by relative risk (RR) and its corresponding 95% confidence interval (CI). Fixed or random effect model was selected based on the results of heterogeneity test. Further, subgroup analyses were conducted to explore the association of 5-HTTLPR polymorphism and autism risk in different population.Results: Eleven studies with 930 cases and 1234 controls were identified. Although there was a significant association between 5-HTTLPR polymorphism and autism risk under the dominant model after removing the studies causing heterogeneity, the significance did not exist after Bonferroni's correction. Subgroup analyses also showed similar results after Bonferroni's correction. In addition, there was no obvious publication bias in our meta-analysis.Conclusions: Our present meta-analysis does not support a direct effect of 5-HTTLPR polymorphism on autism risk according to present results. Further analyses of the effect of genetic networks and more well designed studies with larger sample size are required. Autism spectrum disorder (ASD) is defined as a group of neurodevelopmental disorders classically characterized by impaired social communication, repetitive behaviors and selective attention , 2. In t+/Cl− dependent solute-carrier family of transporters (NG_011747.2), which is located on chromosome 17q11.1-17q12, containing 15 exons. The serotonin-transporter-linked promoter region (5-HTTLPR), located in the promoter region of the SLC6A4 gene, has been widely studied for its role in altering 5-HTT messenger RNA (mRNA) levels. A functional insertion-deletion polymorphism in the promoter region gives rise to two main alleles, the short (S) (14 repeats) and long (L) (16 repeats) variants  is one of the most common members in monoaminergic signaling system and belongs to the Nasporters . It coulvariants –15. The long (L)  repeats ers NG_0147.2, whiic cleft –20.In 2008, Huang et al. conducted a meta-analysis using family-based studies and showed preferential transmission of the s allele of 5-HTTLPR in US mixed population samples . NeverthTo identify the relevant records which explored the association of 5-HTTLPR polymorphism with autism risk, we performed a systematic literature search of PubMed, Embase, Web of Knowledge and CNKI  databases up to June 27, 2018. The combination of terms “SERT,” “5-HTT,” “SLC6A4” or “serotonin transporter gene”; “autism”, “autistic” or “ASD”; and “polymorphism,” “variation,” “mutation,” or “genotype” was used in order to obtain all the genetic studies on the association of 5-HTTLPR genetic polymorphism and autism. The references of retrieved studies were also examined to identify additional studies. For the related data not revealed in original articles, we tried to contact their authors.In our meta-analysis, records were included using the following criteria: (1) case-control studies evaluating the association between 5-HTTLPR polymorphism and autism risk; (2) the number or frequency of genotypes was provided in studies or obtained by contacting the authors; (3) records were clinical studies. The exclusion criteria were: (1) studies with insufficient information such as genotype frequency or number not obtained were excluded; (2) transmission/disequilibrium test (TDT) studies; (3) review, comment or abstract; (4) patients with genetic disorders associated with ASD were excluded. If the same population was included in previous studies, only the one with the largest sample size was included.Data extraction was conducted independently by two investigators in order to reduce personal error. The following information from each study was recorded: first author's name, publication year, country, ethnicity, diagnostic criteria, age group, genotyping method, genotype number for cases and controls. A standardized data collection protocol was used for information collection. Discrepancies were settled by discussion and consensus.I2 value and Chi-square based Q-test. A significant Q statistic (P < 0.10) or I2 > 50% indicated heterogeneity across studies and a random effect model was used to calculate corresponding RRs; otherwise, a fixed effect model was selected. Z test was used to determine the statistical significance of RRs. In addition to overall analysis, subgroup analysis was conducted in Chinese Han population and children. To assess the stability of our results, we performed sensitivity analysis by removing each study in turn. Hardy-Weinberg equilibrium was tested based on the reported genotype frequencies among controls. To evaluate whether publication bias was a threat to the validity of our results, Begg's funnel plot  and Egger's test were applied. At last, Bonferroni's correction was conducted to control the type I error rate in our meta-analysis. All statistical tests were performed using STATA 12.0 software .We performed the meta-analysis with the allele comparison model (s vs. l), the codominant model (ss vs. ll), the recessive model (ss vs. ls/ll) and the dominant model (ss/ls vs. ll) for the unknown inherited model of autism. The pooled relative risk (RR) and its corresponding 95% confidence interval (CI) were used to evaluate the strength of association between 5-HTTLPR polymorphism and autism risk. Fixed effect model or random effect model was selected based on heterogeneity analysis, which was assessed by A total of 853 records were identified update to June 27, 2018 . Two hundred and seventy-two duplications were found in the records and 544 records were excluded for the unmatched titles or abstracts. Full text reading help removed 26 records  from the last 37 records. At last, 11 records (11 studies) –33 with The characteristics of included studies were listed in RR = 1.13, 95%CI 0.95–1.34, P = 0.163; for ss vs. ll: RR = 1.20, 95%CI 0.82–1.78, P = 0.351; for ss vs. ls/ll: RR = 1.08, 95%CI 0.73–1.58, P = 0.704; for ss/ls vs. ll: RR = 1.11, 95%CI 0.91–1.35, P = 0.289;) .In our present meta-analysis, eleven studies with 930 cases and 1,234 controls determined the effect of 5-HTTLPR l/s polymorphism on autism risk. Results were shown in  0.289;) . Neverth 0.289;) . HoweverI2 value and Chi-square based Q-test to evaluate heterogeneity across selected studies. Obvious heterogeneity was observed in overall comparisons under all genetic models (P = 0.768).We employed c models . Thus, rc models , 27, 31;c models , 27, 30;c models ; ss/ls vc models , 31, 32]c models . After rc models . NeverthTo examine the influence of each study on the overall RRs, we carried out sensitivity analysis by excluding one individual study from the overall pooled analysis sequentially. Our results of sensitivity analysis revealed that the pooled RRs and its corresponding 95% CIs were generally similar when we removed any study under all the four genetic models. Hence, our results were relatively stable and credible.P = 0.212; for ss vs. ll: P = 0.835; for ss vs. ls/ll: P = 0.736; for ss/ls vs. ll: P = 0.294).Begg's funnel plot and Egger's test were conducted to assess the publication bias qualitatively and quantitatively. Although the shape of the funnel plots was slight asymmetry , the resIn the past few decades, epidemiological studies have shown that environmental factors, such as air pollution, heavy metals, pesticides, psychological stress and nutritional factors, have an impact on the development of ASD . In addiOur meta-analysis represents a comprehensive and systematic assessment of the relation of 5-HTTLPR polymorphism with autism risk. So far, there are about eleven population based studies reported to evaluate the association between 5-HTTLPR polymorphism and autism risk. However, the results were inconsistent. A previous meta-analysis conducted by Huang et al. only explored the effect of this polymorphism based on families in 2008 . CompareIn our present meta-analysis, although the pooled RRs and 95%CIs under different comparison models did not show strongly significant association of 5-HTTLPR polymorphism with autism risk in overall populations under four genetic models, heterogeneity across studies may have a significant impact on results. Galbraith plots showed some studies may cause heterogeneity. After removing those studies, we found an obvious association between 5-HTTLPR polymorphism and autism risk under the dominant model. However, the association was not found under the dominant model after Bonferroni's correction. In addition, subgroup analyses on children and Chinese Han also exhibited the similar results under the dominant model. Compared with previous results, it may be caused by the limitation of population. Therefore, we hypothesized that 5-HTTLPR polymorphism may not have a direct effect on the risk of autism, which was in consistent with some recent GWAS studies. Taken together, more studies are needed to evaluate the effect of this polymorphism on autism.Some limitations in our meta-analysis should be pointed out. Firstly, the number of studies and samples in our meta-analysis was limited for the insufficient information of some studies and unpublished records, which may influence the synthesized RRs. Secondly, autism was a complex disease and many factors may contribute to the development of it. We could only discuss the factor of ethnicity for the limited information of studies. Thirdly, some TDT studies were not included in our study to eliminate family factors. Fourthly, although we find no obvious difference in allele frequencies across ethnic groups based on present information, studies with more information about population, especially Chinese Han, should be conducted to further demonstrate our conclusion. Finally, the lack of information about the tri-allelic model of gene made us fail to analyze our data by using a tri-allelic model.In conclusion, the present study suggested that the 5-HTTLPR polymorphism may not be associated with the risk of autism. Further analysis of studies based on larger sample size and homogeneous autism patients is required to explore the association between the 5-HTTLPR polymorphism and autism risk.JG and XF conceived the study. HW and FY collected the data and drafted the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


The melatonin receptor 1B (MTNR1B) polymorphism rs10830963 C>G has been reported to be associated with the risk of gestational diabetes mellitus (GDM) with inconsistent results. To clarify the effect of the polymorphism on the risk of GDM, a meta-analysis therefore was performed. Pooled OR with its corresponding 95%CI was used to estimate the strength of the association. Totally 14 eligible studies with a number of 5033 GDM patients and 5614 controls were included in this meta-analysis. Results indicated that the variant G allele was significantly associated with an increased GDM risk . In the stratified analysis by ethnicity, similar results were found in Asians  and in Caucasians . FPRP and TSA analyses confirmed findings support that the rs10830963 G allele increases the risk of GDM, and further functional experimental studies are warranted to explore and clarify the potential mechanism. Gestational diabetes mellitus (GDM) is defined as abnormal glucose tolerance with onset or first recognition during pregnancy . WorldwiMTNR1B is located on human chromosome 11q21-q22, spanning about 22 kb and consisting of 3 exons and 1 intron. So far, 64 single-nucleotide polymorphisms (SNPs) have been validated in the MTNR1B gene (http://www.ncbi.nlm.nih.gov/SNP), and some of which were reported to be associated with GDM risk. The SNP rs10830963 is located in the unique intron between exon 1 (+5.6 kb) and exon 2 (−5.9 kb) of MTNR1B gene. Genotype–phenotype study of the SNP rs10830963 C>G showed that compared with the wild-type C allele of rs10830963, the variant G allele was associated with increased MTNR1B transcript levels in human islets [n islets . The stuMTNR1B rs10830963 C>G on the risk of GDM, we therefore performed a meta-analysis with a total of 5033 GDM patients and 5614 controls from 14 published case–control studies.Because of the functional consequence of rs10830963 C>G, many association studies have examined its effect on the risk of GDM . HoweverMTNR1B rs10830963 polymorphism with the risk of GDM. The following key words ‘MTNR1B’ or ‘Melatonin receptor 1B’, ‘gestational diabetes mellitus’ and ‘variation’ or ‘polymorphism’ were used. The corresponding Chinese terms were used in the Chinese library. All studies were published up to March 1, 2019. In addition, we manually searched for additional published studies on this topic in the references cited in the retrieved studies.We searched NCBI PubMed, Google Scholar and the Chinese National Knowledge Infrastructure (CNKI) databases for the association studies of MTNR1B rs10830963 C>G polymorphism and GDM risk; case–control study; genotype or allele distribution information in cases and controls for calculating odds ratio (OR) with corresponding 95% confidence interval (CI); the study was written in English or Chinese. Accordingly, family based studies, abstract, case reports, comments and reviews were excluded. If studies had overlapped subjects, only the largest study was included in the final analysis.The included studies in this meta-analysis had to meet the following criteria: evaluation of the P-value of the Chi-square goodness of fit test for Hardy–Weinberg equilibrium (HWE) in controls.Two professional investigators (Huang B and Wang Y) independently reviewed the articles and extracted the data from eligible publications. The following information was extracted from each included study: first author, year of publication, country, diagnostic criteria, source of controls, number of cases and controls, genotype or allele distribution data of cases and controls, mean age, mean body mass index (BMI) and In the present study, the Newcastle–Ottawa scale (NOS) scale was used to assess the quality of the eligible studies by Yu, X.Y. and Wang, Y.K. independently . A scoreMTNR1B rs10830963 C>G polymorphism in controls from HWE was tested by using the Chi-square goodness of fit test, and a P-value less than 0.05 was considered a departure from HWE. The odds ratio (OR) and 95% confidence interval (CI) were used to assess the strength of association between the MTNR1B rs10830963 C>G and GDM risk. The heterogeneity across the studies was assessed by the Q test, and was considered significant when a P-value less than 0.1 [P-value < 0.05 was considered as an indication for the potential presence of publication bias. Sensitivity analyses were done to assess the influence of individual study on the pooled ORs. All analyses were performed by using Stata software, version 12.0 . In addition, false positive report probability (FPRP) was estimated to assess the robustness of foundings statistically significant association by using the method described by Wacholder et al. [P-value. An FPRP less than 0.2 was considered as a noteworthy association [Deviation of genotype frequencies of the than 0.1 . A fixedthan 0.1 . The potthan 0.1 ,40. A P-r et al. . The FPRociation ,42.Z-curve crosses the TSA boundary or enters the insignificance area, a sufficient level of evidence has been reached, and no further studies are necessary. However, when the Z-curve does not exceed any of the boundaries and the required sample size has not been reached, evidence to reach a conclusion is insufficient [Meta-analysis might be affected by type I error due to the increased risk of random error and repeated significance testing . Trial sfficient . Review MTNR1B rs10830963 C>G and GDM risk. Finally, 14 studies with 5033 GDM patients and 5614 controls were selected in present study. According to the evaluation of NOS scale, 12 of the included literatures were considered to be of high quality (score ≥ 5) and 2 of them were of low quality (score < 5). Main characteristics of the included studies are shown in The flowchart of study selection for this meta-analysis is presented in P < 0.001, Pheterogeneity = 0.090, 2I = 35.6%; GG vs. CC: OR = 1.78, 95% CI = 1.45−2.19, P < 0.001, Pheterogeneity = 0.001, 2I = 61.9%. G vs. C: OR = 1.33, 95% CI = 1.21−1.47, P < 0.001, Pheterogeneity = 0.001, 2I = 64.1%), respectively.As shown in P = 0.020, Pheterogeneity = 0.633, 2I = 0.0%; GG vs. CC: OR = 1.52, 95% CI = 1.23−1.89, P < 0.001, Pheterogeneity = 0.069, 2I = 48.7%; G vs. C: OR = 1.23, 95% CI = 1.10−1.37, P < 0.001, Pheterogeneity = 0.064, 2I = 49.7%) and in Caucasians .In the stratified analysis by ethnicity, as shown in Q test was used to evaluate the heterogeneity across the included studies and the heterogeneity across studies was found in most of comparisons. As shown in Q test suggested that a low to high heterogeneity across studies presented in most of comparisons. We then used the meta-regression analysis to explore the source of heterogeneity by ethnicity (Asian and Caucasian), diagnostic criteria (ADA and others) and assessed literature quality (score ≥ 5 and <5), and found that they didn’t contribute to the main observed heterogeneity across the studies, effect of ethnicity , diagnostic criteria  and literature quality .In this present study, the MTNR1B rs10830963 C>G polymorphism and GDM risk were performed to assess the stability of the pooled ORs under the CG vs. CC and GG vs. CC comparisons. The leave-one-out analysis showed that no single study dramatically influenced the pooled ORs .Both Begg’s and Egger’s tests were performed to evaluate the publication bias of the included studies. The shape of the funnel plots did not reveal any evidence of obvious asymmetry for all genetic models  and 8, aMTNR1B rs10830963 C>G polymorphism and GDM risk. At the prior probability of 0.1 and FPRP cut-off value of 0.2, the FPRP values for the significant findings in the heterozygous genotype comparison (CG vs. CC), homozygote model (GG vs. CC) and allele model (G vs. C) of overall were 0.005, 0.008 and 0.004, respectively. Moreover, the FPRP values for the significant findings in the studied three models were 0.153, 0.009 and 0.010 in Asians, 0.007, 0.047 and 0.006 in Caucasians, respectively. As shown in FPRP was adopted to assess the noteworthiness of the significant associations between the TSA was performed to reduce the random errors and increase the robustness of the conclusions. The TSA of the heterozygote (CG vs. CC) and homozygote models (GG vs. CC) for rs10830963 C>G among Asians and Caucasians –12, indiGestational diabetes mellitus (GDM) is considered as an early form of diabetes, which might increase the risk of adverse pregnancy outcomes and substantial long-term adverse health among mothers and their offspring. Pregnant women with GDM are at a relatively high risk of developing Type 2 diabetes in the future . MoreoveMTNR1B rs10830963 G allele was relatively higher in GDM cases than that in controls, suggesting a significant association with an increased GDM risk.A number of candidate gene-based association studies and genome-wide association studies (GWAS) have shown that some GDM relation genetic variants could provide insight into pathogenetic mechanisms underlying the disease . Functio. revealed a significant association between MTNR1B and the glucose disposition index and acute insulin response among T2DM [MTNR1B and to be related to the risk of T2DM or GDM [MTNR1B rs10830963 C>G might possibly increase the incidence of GDM. This may due to the observation that the increased expression of MTNR1B on β cells diminished intracellular cyclic cAMP level, thereby inhibited the glucose-stimulated insulin secretion.GDM could occur when the pancreatic islet β cells were impaired by an increased insulin resistance during pregnancy . MTNR1B ong T2DM . MeanwhiM or GDM . Our stu. observed a statistically significant association between the MTNR1B variant CG/GG genotype and GDM risk  [MTNR1B rs10830963 C>G polymorphism on GDM risk in different ethnic populations.We summarized the evidence to date with 5033 GDM patients and 5614 controls by a meta-analysis with TSA, and found that compared with the wild CC genotype, the variant CG and GG genotype were significantly associated with an increased risk of GDM, respectively. Furthermore, the subgroup analysis by ethnicity revealed that rs10830963 C>G polymorphism was significantly associated with the risk of GDM both in Asians and in Caucasians. Obvious heterogeneity across studies was observed in data processing in this meta-analysis. We then used the meta-regression analysis to explore the potential source of heterogeneity across the studies by ethnicity, diagnostic criteria and assessed literature quality, and found that they didn’t contribute to the main observed heterogeneity.A previous meta-analysis by Zhang et al14−1.35) . HoweverMTNR1B rs10830963 C>G and GDM risk was much lower than the preset threshold. It suggests that the positive findings both in overall analysis and racial-related subgroup analysis of heterozygous and homozygous models are probability authentic and reliable. Hence, we believe the association of rs10830963 C>G and GDM risk are credible to some extent. Further, a TSA indicated that the cumulative Z-curve crossed the conventional cut-off value and the TSA boundaries both in Aisan and Caucasian subgroup analyzes and confirmed that the rs10830963 SNP was mainly associated with susceptibility to GDM. The results of TSA suggesting that no additional researches are required to further evaluate the findings.FPRP analysis is an efective approach to verify the noteworthiness of signifcant association fndings. In the present study, we performed a relatively stringent FPRP threshold of 0.2. We found that the FPRP values of the observed signifcant associations between MTNR1B rs10830963 and GDM risk. However, some limitations should be point out in the current meta-analysis. First, although the Begg’s and Egger’s tests did not detect any significantly statistical evidence of publication bias, selection bias could exist because only published case–control studies were included. Second, the small sample size might limit the statistical power of the study. Third, the findings of present study were based on the unadjusted results. Due to lack of individual-level data prevented us from making further analysis to identify any genotype–environment interaction between rs10830963 C>G and metabolic traits, such as FPG, pancreatic β-cell function, acute insulin response or indices for insulin sensitivity.The current meta-analysis more comprehensively makes the relationship clear between MTNR1B rs10830963 C>G polymorphism significantly increases the risk of GDM. Further functional experimental studies are warranted to help explore and clarify the potential mechanism.In summary, the current meta-analysis with TSA indicates that the variant G allele of


No significant associations were observed with DLBCL, FL, or MZL. Our findings suggest that there may be some shared genetic factors between CLL and height, but other endogenous or environmental factors may underlie reported epidemiologic height associations with other subtypes.Although the evidence is not consistent, epidemiologic studies have suggested that taller adult height may be associated with an increased risk of some non-Hodgkin lymphoma (NHL) subtypes. Height is largely determined by genetic factors, but how these genetic factors may contribute to NHL risk is unknown. We investigated the relationship between genetic determinants of height and NHL risk using data from eight genome-wide association studies (GWAS) comprising 10,629 NHL cases, including 3,857 diffuse large B-cell lymphoma (DLBCL), 2,847 follicular lymphoma (FL), 3,100 chronic lymphocytic leukemia (CLL), and 825 marginal zone lymphoma (MZL) cases, and 9,505 controls of European ancestry. We evaluated genetically predicted height by constructing polygenic risk scores using 833 height-associated SNPs. We used logistic regression to estimate odds ratios (OR) and 95% confidence intervals (CI) for association between genetically determined height and the risk of four NHL subtypes in each GWAS and then used fixed-effect meta-analysis to combine subtype results across studies. We found suggestive evidence between taller genetically determined height and increased CLL risk (OR = 1.08, 95% CI = 1.00–1.17,  In epidemiologic studies, taller adult height has been associated with an increased risk of several subtypes of non-Hodgkin lymphoma (NHL) ; howeverThe reported positive associations between height and multiple subtypes of NHL suggest the existence of common biological pathways, possibly mediated by genetic factors. Height is a highly polygenic trait and estimated to be 70–80% heritable based on twin studies –19. GenoIn this study, we used previously established height-related genetic variants to construct polygenic risk scores (PRS) for height in order to examine the associations between genetically-inferred height and the risk of four common NHL subtypes: DLBCL, FL, CLL, and MZL.We used data from eight GWAS of NHL subtypes in populations of European ancestry , the detGenotyping was performed on commercially available Illumina and Affymetrix platforms . Standarp-values for Hardy-Weinberg equilibrium >1 × 10−5. To generate the PRS, the height-increasing allelic dosage for each SNP was multiplied by the reported meta-analysis beta coefficient  at the GWAS level. Since all controls in the UCSF1/NHS GWAS were female, models for this GWAS were adjusted for age and significant principal components of ancestry only. In a sensitivity analysis, we also adjusted for genetically determined body mass index (BMI), using previously established loci. The regression analyses were performed in R using the glm function.As previous studies have shown etiologic heterogeneity among NHL subtypes , we analI2 metric . The UCSF1/NHS GWAS was excluded from the sex-specific meta-analysis of FL because all controls were female. Between-sex heterogeneity was evaluated by performing a fixed-effect meta-analysis to obtain the p-value from the Cochran's Q-test and I2.For each NHL subtype, GWAS-specific risk estimates were pooled using fixed-effect meta-analysis . Only on2 metric . Meta-anp = 0.049), with the PRS modeled as a continuous variable (p = 0.04) than among males , but the between-sex heterogeneity was not statistically significant (p = 0.34). No significant heterogeneity among studies was observed for CLL among males and females combined . With regards to sex-specific analyses for these three subtypes, no significant associations were seen among men or women, with risk estimates similar to those observed for both sexes combined.No association was observed between genetically predicted height and the risk of DLBCL , FL , or MZL  . AdjustmIn this first examination of genetically inferred height and NHL risk, we observed a borderline positive association between genetically inferred taller height and the risk of CLL. This finding is consistent with observations from epidemiologic studies examining height and CLL risk, which generally have reported positive or insignificant associations , 5, 11. No evidence of association was observed between genetically determined height and the risk of DLBCL, FL, nor MZL. Although the findings are not consistent , 8, someDespite the suggestive associations in the literature for self-reported or measured height and FL and DLBCL risk, there are several possible reasons why our study of genetically determined adult height did not identify statistically significant associations. A recent methodologic paper on the use of polygenic risk scores demonstrated that null results are often explained by inadequate sample size . Even thAmong the noteworthy limitations of our study is the fact that our participants were from populations of European descent, limiting the generalizability of results. However, this restriction to Europeans also has the benefit of minimizing bias from population stratification, which is an important concern with studies of height. We did not adjust for multiple testing, because previous studies have shown individual NHL subtypes to have distinct etiologies . HoweverIn conclusion, we present the first polygenic risk score analysis examining the genetic contribution of attained adult height on the risk for the four most common NHL subtypes. Although we did not observe statistically significant associations with DLBCL, FL, or MZL, we observed evidence of a positive association between genetically-inferred height and CLL risk, lending more support to previous observational studies. Additional studies are warranted to explore the association between genetically-inferred height and risk of other NHL subtypes, particularly those that have been reported to be associated with self-reported or measured height in epidemiologic studies , 35. Addhttps://www.ncbi.nlm.nih.gov/projects/gap/cgi-bin/study.cgi?study_id=phs000801.v2.p1.The datasets generated for this study can be found in the dbGap phs000801.v2.p1 at All studies involving human participants were reviewed and approved by their respective Institutional Review Boards. The patients/participants provided their written informed consent to participate in this study.AMoo and SIB organized and designed the study. AMoo, EK, LRT, MJM, OP, AMon, NW, ZW, CFS, JRC, and SIB contributed to the design and execution of data analysis. AMoo and SIB wrote the first draft of the manuscript. EK, LRT, AMon, NW, CFS, SLS, RV, CV, YZ, KS, JS, VJ, GG, BL, AA, AN, PMB, NJC, GS, WC, HH, H-OA, DA, NB, YB, SB, PBo, PBr, AB-W, FC, JC, LC, DC, KC, GF, LF, SG, HG, MG, BG, TH, RJ, QL, ML, MMa, JM, MMe, LM, RM, TM, LMM, KN, KO, MPu, AP, VR, ER, SS, RS, MS, AS, LFT, RT, EW, SW, TZ, NR, BB, JRC, and SIB conducted the epidemiological studies involved this study. All authors contributed to the writing of the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.


TNF-α) plays an important role inautoimmune diseases. Previous studies have investigated the association ofTNF-α-238G/A (rs361525) and -308G/A(rs1800629) polymorphisms with rheumatoid arthritis (RA) and systemic lupuserythematosus (SLE). However, no agreed conclusion had been made. Therefore,this meta-analysis was conducted to assess the associations ofTNF-α-238G/A and -308G/A polymorphismswith RA and SLE risk. A systematic search was conducted in commonly useddatabases. Meta-analysis was performed by STATA12.0. A total of 43 studies wereincluded. In the overall population, the TNF-α-238A allele wasobserved to be a protective factor for RA  and the TNF-α-308A allele was foundto be a risk factor for SLE . However, no evidence of association was found betweenTNF-α-238 G/A polymorphism and SLE nor between -308G/A andRA. In the subgroup analysis, TNF-α-308A allele played apathogenic role for RA in Latin Americans  and for SLE in Latin Americans  and Europeans , while it played a protective role for RAin Asians . Nosignificant association was found between TNF-α-308G/A and SLEsusceptibility in Africans and Asians. This meta-analysis demonstrated thatTNF-α-238A was associated with decreased risk of RA ratherthan SLE, while -308G/A polymorphism was associated with SLErather than RA. Stratification analysis indicated that different ethnicitieswould have different risk alleles.Tumor necrosis factor-alpha ( Rheumatoid arthritis (RA) and systemic lupus erythematosus (SLE) are both chronicsystemic autoimmune diseases. RA is associated with systemic and chronicinflammation of the joints, resulting in synovitis and pannus formation. It is aserious pathological condition that can lead to incapacitation and decreased lifeexpectancy compared with the normal population . People TNF-α gene, which is located inchromosome 6, within the class III region of MHC  and position-308 , attracted widespread attention.Several case-control studies were conducted to investigate the association ofTNF-α-238G/A and -308G/A with RA and SLE.However, significant associations with these two polymorphisms related to RA and SLEwere not consistently observed. Some studies indicated a significant association,abnormalities have been reported in RA and SLE ,7. TNF-αn of MHC . Several and SLE –12 andociation–15 whilociation–18. The ociation,20. Thesn Asians ,16,17,21f RA/SLE,23. It wTNF-α”,“polymorphism”, “variant”, “rheumatoid arthritis”, “RA”, “systemic lupuserythematosus”, and “SLE” were searched as both medical subject heading (MeSH)terms and text words. In addition, the reference citations in the obtainedarticles were scrutinized to guarantee not missing eligible studies.A comprehensive search was conducted until July 31, 2018 in Medline, Embase,Chinese Biomedical Literature Database (CBM), and China National KnowledgeInfrastructure (CNKI). “Tumor necrosis factor-alpha”, “a) be a case-control study andbased on original data (independence among studies); b) assessthe presence of the variant TNF-α-238G/A and/orTNF-α-308G/A in the TNF-α gene of patientswith and without RA or SLE; c) provide sufficient data tocalculate odds ratio (OR). We excluded the following: a) nocontrol group; b) studies containing overlapping data ; c) no certain number of thenull and/or wild genotypes.For inclusion, a study had to: The following information was extracted from the studies: first author's name,publication time, country, ethnicity, study design, sequencing method, number ofcases (RA or SLE patients) and controls , mean age, femaleproportion, genotype distribution, and mutation sites. All data were collectedby two members independently. The other investigators were consulted to reach aconsensus when any divergence occurred.The quality assessment of the included studies were conducted by two membersindependently according to the quality assessment scale . In thisvs GG), dominantmodel (GA+AA vs GG), and allelic comparison (Avs G). The overall effect was appraised through the Z test,which could be deemed significant if the P value was less than 0.05. Theheterogeneity for the included articles was evaluated with Cochran's Q test,I2 statistics (the heterogeneity could be accepted if P>0.1and I2 ≤50%)  . If theion bias . BesidesTNF-α and 12 articles were without allele or genotypingdata, 7 articles were not case-control studies, 2 articles were with overlappingdata (reference not shown). Another 5 studies were excluded due to the absenceof Hardy-Weinberg equilibrium (reference not shown). Finally, a total number of43 relevant studies that met the inclusion criteria were included in thismeta-analysis, among which, 14 case-control studies focused onTNF-α-238G/A and 42 studies focused onTNF-α-308G/A.The flow chart  describiTNF-α-238G/A andTNF-α-308G/A were presented separately, each of them wastreated as separate studies. In the end, in terms ofTNF-α-238G/A, 8 studies containing 1386 cases and 1535controls for RA and 7 studies involving 1296 cases and 1558 controls for SLEwere included in this meta-analysis. In terms of TNF-α-308G/A,19 studies including 3503 cases and 3993 controls for RA and 26 studiesinvolving 3051 cases and 4232 controls for SLE were included in thismeta-analysis.In the studies whose genotype frequencies of TNF-α-238G/A polymorphism with RA is shown in 2 statistic except in European alleliccomparison. Therefore, fixed-effects model was used except for the Europeanstudy. Overall, an association was found in every genetic model. However,stratification by ethnicity indicated that TNF-α-238G/A was notsignificantly associated with RA in Asians, while an association was found inLatin Americans  and -238G/A (rs361525) of transcriptionstart site, respectively  in human B cell line, which may be related withthe generation of a hypersensitive site at position -308 and an adjacent area ofprotection (TNF-α-238A in TNF-α transcription wasuncertain since some studies indicated that A allele or GA genotype affected geneexpression while other studies did not (TNF-a may have aneffect on the susceptibility to autoimmune disease development and on its clinicalmanifestations.The premise of this meta-analysis of ividuals . Meanwhicription,37. TNF- did not ,40. TNF-TNF-α gene and the susceptibility to RA and SLE. In addition,compared with the former meta-analysis about the association betweenTNF-α-308 and RA (Several strengths characterize this meta-analysis. Firstly, it is the firstmeta-analysis that focused on the association between two SNPs of the8 and RA ,more stHowever, this study has several potential limitations. Firstly, a large amount ofstudies was included in this meta-analysis and inevitably resulted in heterogeneity,which could not be eliminated easily. Random-effects model was selected to reducethe potential bias. Additionally, the sensitivity analysis was conducted and the ORswere not significantly affected. Therefore, the influence on the accuracy of ourresults was limited. Secondly, when taking subgroup analysis, a small number ofstudies were eligible and the evidence seemed to be insufficient and unconvincing.Finally, only the data of articles published in English or Chinese were extractedand a potential bias might thus have been introduced. The findings in this studyshould be interpreted prudently considering these limitations.TNF-α gene in RA or SLE development. Further evaluation of theeffect of gene-gene and gene-environment interactions on the TNF-αpromoter-238G/A and -308G/A polymorphism and RA/SLE susceptibilityis necessary. Moreover, in terms of variants in TNF-α at position-238 or -308, whether and how this change promotes the inflammation process of RA orSLE requires further investigation.The results of stratification analysis still need more large-scale, well-designedcase-control studies to be proven. However, our study provided an argument forperforming further mechanistic studies to better understand the role of the


CDKN2A were considered to play an important role on leukemogenesis. Two single nucleotide polymorphisms (SNPs) at CDKN2A locus were identified to impact on ALL susceptibility via genome wide association studies, and followed by multiple subsequent replication studies at the specific hits. Here, we conducted a systematic review and meta-analysis to re-evaluate the association of both SNPs (rs3731217 and rs3731249) with ALL susceptibility by gathering the data from 24 independent studies, totally containing 7922 cases/21503 controls for rs3731217 and 6295 cases/24191 controls for rs3731249. Both SNPs were significantly associated with ALL risk (odds ratio [OR] = 0.72 and 2.26 respectively), however, exhibit race-specific pattern. In summary, our meta-analysis indicated that two SNPs at CDKN2A locus are associated with ALL susceptibility independently mainly in Caucasians. Future large-scale studies are required to validate the associations in other ethnicities.Acute lymphoblastic leukemia (ALL) is the most common cancer in children, and alterations in  CDKN2A gene, which encodes two important tumor proteins (p14ARF and p16INK4A), is commonly observed in leukocytes of ALL patients /Q100%, varying from 0 to 100%. Generally, I2 = 50% is a threshold value, when I2 < 50% and P value >0.1, heterogeneity among studies were acceptable and fixed-effect model is more suitable to compute the merge OR and 95% CI. On the contrary, if I2 > 50% and P value <0.1, indicating highly heterogeneity, were existed in those studies and random-effect model was utilized to compute the merge OR and 95% CI. When necessary, subgroup analyses can be employed.Heterogeneity among studies was assessed through tatistic , where QSensitivity test carried out by omitting each of the studies discussed the association of rs3731217 (or rs3731249) with ALL susceptibility, pooled OR and 95% CI are not of significant difference (Supplementary Tables S5 and S6), which in turn certified the robustness of the relationship between rs3731217 or rs3731249 and ALL predisposition.http://www.ohri.ca/programs/clinical_epidemiology/oxford.asp). The scales evaluated study qualities in several aspects (Supplementary Table S10): (1) inclusion and exclusion criteria of patients, (2) source of controls, (3) comparability of cases and controls, (4) sample size, (5) quality control of genotyping methods, and (6) HWE. Strictly obey the principles, two investigators separately evaluated and scored each study and the final results reached an agreement with a third author participated in the discussion. The final score showed in Supplementary Tables S11 and S12, and the higher the score, the better is the quality of the study.The quality estimation of the researches was based on the methodological quality assessment scale, which was adjusted from the Newcastle–Ottawa assessment scale for case–control studies  by Begg’s funnel plot and Egger’s test . Further, HWE was assessed by Chi-square test among the included studies in both SNPs studies respectively (Supplementary Tables S8 and S9).CDKN2A locus. For rs3731217, association of this SNP with ALL susceptibility was first reported in 2010 through GWAS approach, and investigated in other three GWASs or follow-up candidate studies. For the missense SNP rs3731249 (A148T), its influence on ALL risk was first reported in 2015, and was validated in another three studies. Since rs3731249 and rs36228834 are in perfect linage disequilibrium (LD) with each other (r2 = 1) in Caucasians, an earlier study can thus be included in the meta-analysis for rs3731249. The characteristics of 19 studies on rs3731217 and 10 studies on rs3731249 were summarized in Through literature search with keywords (see Methods), 24 independent researches manifested in 16 literatures met the inclusion criteria, and were selected for meta-analyses , describP=0.87 and I2 = 0%, P<0.00001, P=0.51 and I2 = 0%) and recessive model (P=0.86 and I2 = 0%) among 12 out of 19 studies that individual genotypes can be got . In addition, no heterogeneity was observed in the dominate model (P=0.19 and I2 = 32%) and recessive model (P=0.68 and I2 = 0%) among 7 out of 10 studies that individual genotypes can be got . The result of sensitivity analysis showed that the association between rs3731217 (or rs3731249) and ALL risk doesn’t significantly fluctuate when removing each of the studies (Supplementary Tables S5 and S6).r2<0.01 in Caucasians and Hispanics based on 1000 genome data , which was consistent with the fact that these two SNPs associated with ALL risk independently in conditional model.Variant allele of rs3731217 was common across all ethnicities, while variant allele of rs371249 was only common in Caucasians and Hispanics, rare in Africans, even not observed in Asians. It could be the reason that no association between rs3731217 with ALL risk if rs3731249 was tagged by this SNP, we next investigated the genetic characteristic of CDKN2A locus across ethnicities, and identified the similar LD pattern between Caucasians and Hispanics, much less extensive in Africans and Asians. However, rs3731217 and rs3731249 were not in the same LD region, with INK4A, protein product of CDKN2A, and reduced the tumor suppression effect of p16INK4A, suggesting rs3731249 act as a causal variant for ALL susceptibility. We next investigated the potential causal variant tagged by rs3731217 by using webtool (Haploreg), because association of this SNP with ALL can’t explained by rs3731249. Totally 17 known SNPs showed moderate LD (r2 > 0.4) with rs3731217 in Caucasians (Chr9:21956078-22039426) (Supplementary Table S7). Among these SNPs, rs2811711 is located within the region marked by activated promoter or enhancer histones in multiple tissue types, especially in B cells, based on the chromatin state information from Roadmap and ENCODE database , whereas inconsistent associations were noticed in others (e.g. CDKN2A). Considering CDKN2A acts as an important tumor suppressor gene and is frequently inactivated in somatic leukemic cells [N = 4254), rare in Africans , common in Hispanics  and Caucasians  according to the largest exome database  [CDKN2A SNPs in Asian population raises a possibility that these two SNPs are in the same LD region, and can be explained by the same causal variant. However, rs3731249 poorly links to rs3731217 across all ethnicities (r2 < 0.01) , and is associated with ALL susceptibility independently according to conditional analysis, suggesting multiple causal variants located within the CDKN2A locus. Moreover, we combined the recent studies and noticed that wild-type allele of rs3731249 is tend to be lost through either loss of heterozygosity or post-transcriptional inactivation during the leukemogenesis process significantly.A series of GWAS approaches had identified at least several loci those are significantly associated with ALL susceptibility. After the extensive replication studies, some loci can be validated in all independent multiethnic cohorts  SNPs associated with >1000 traits (http://www.ebi.ac.uk/gwas/) [INK4A. Interestingly, we and another independent group also observed that the risk allele is enriched in the leukemic cells through either loss of heterozygosity in DNA level or allele-specific epigenetic modification [BCL11A, which altered the GATA1 and TAL binding motifs of BCL11A enhancer region respectively [CDKN2A exon 3 [By the end of 2017, ∼2110 GWASs had published, reporting around 18000 genome-wide significant (k/gwas/) , but onlk/gwas/) , while tk/gwas/) . Great ek/gwas/) , and rs3fication . In anotectively . The sysectively . ActuallA exon 3 , which r


Genetic risk factors can provide insight into susceptibility for acute pancreatitis (AP) and disease progression towards (infected) necrotizing pancreatitis and persistent organ failure. The aim of the study was to undertake a systematic review of the genetic evidence for AP.Online databases  were searched to 8 February 2018. Studies that reported on genetic associations with AP susceptibility, severity and/or complications were eligible for inclusion. Meta‐analyses were performed of variants that were reported by at least two data sources. Venice criteria and Bayesian false‐discovery probability were applied to assess credibility.SPINK1 (odds ratio (OR) 2·87, 95 per cent c.i. 1·89 to 4·34), IL1B  and IL6  and disease risk. In addition, two novel credible single‐nucleotide polymorphisms were identified in Asian populations: ALDH2  and IL18 . Associations of variants in TNF, GSTP1 and CXCL8 genes with disease severity were identified, but were of low credibility.Ninety‐six studies reporting on 181 variants in 79 genes were identified. In agreement with previous meta‐analyses, credible associations were established for Genetic risk factors in genes related to trypsin activation and innate immunity appear to be associated with susceptibility to and severity of AP. SPINK1, ALDH2, IL1B, IL6 and IL18 with the risk of acute pancreatitis. A lack of replication studies precludes definite conclusions to be drawn regarding the role of genetics in severity and severe clinical phenotypes of acute pancreatitis.A comprehensive systematic review and meta‐analysis of the genetic evidence for acute pancreatitis was conducted, and identified credible associations of Genes affecting trypsin activation and innate immunity important Acute pancreatitis (AP) is a common inflammatory disease of the pancreas with an incidence of 13–45 per 100 000 population in Western countriesBiliary disease and alcohol are the leading causes of AP, and account for 70 per cent of cases; other causes are hypertriglyceridaemia, hypercalcaemia, iatrogenic (mostly related to endoscopic retrograde cholangiopancreatography and drugs) and autoimmune disease, and some are classified as idiopathicPRSS1 gene by Whitcomb and colleaguesSPINK1, CTRC, CASR and CFTRSince the discovery of the causal variant of hereditary pancreatitis in the Field synopsis and meta‐analysis are powerful tools to integrate genetic data from a large field and identify credible genetic associationsAppendix This systematic study was conducted by following a protocol, and is concordant with the Updated Guidance on Human Genome Epidemiology Reviews and Meta‐Analysis of Genetic AssociationsA systematic search of five online databases  was performed, using the keywords and index terms ‘pancreatitis’ combined with ‘mutation*’, ‘polymorphism*’ and ‘variant*’, and applying the limit ‘human’. After removal of duplicates, identified articles were screened for eligibility in two steps: first by title and/or abstract independently by two authors, and second by reading of the full text, and selected based on predefined inclusion and exclusion criteria. The references of identified studies and reviews were cross‐checked for additional studies. In case of disagreement, consensus was resolved by discussion.in vitro and animal studies were excluded. Studies reporting exclusively on paediatric patients (age below 18 years) were also excluded, because of differences in aetiologyGenetic association studies (case–control and cohort) associated with AP and recurrent AP in humans published in a peer‐reviewed journal before 8 February 2018 that met the criteria were included with no restriction on language. Genome‐wide association studies were considered, but not identified. Studies reporting on other than bi‐allelic markers were excluded from quantitative analyses. Studies reporting on tropical, hereditary or autoimmune pancreatitis are considered as chronic pancreatitis, and were excluded. Abstracts, case reports, economic evaluations, Published meta‐analyses of identified variants were considered for comparison. Studies with identical, or largely overlapping, sets of data were compared, and the largest data set was included in the meta‐analysis; when data were identical, the first published study was included.Data extraction was performed separately by two authors, and differences were resolved by discussion. As well as study identifiers such as first author, year of publishing, journal and language, the following demographic data were extracted: study location and design, inclusion and exclusion criteria, sample size, reported outcomes, studied aetiology, ethnic background of patients and control subjects, source of controls, age and sex.The following data points for each variant were extracted: Single Nucleotide Polymorphism database (dbSNP) identifier number, Human Gene Nomenclature Committee symbol and name, type of genetic variation, position and genotype distribution. Genotype counts were calculated when genotype frequencies and numbers of cases and controls were reported. If only allele frequencies were available, genotype counts were calculated assuming distribution according to the Hardy–Weinberg equilibrium (HWE).The corresponding author was contacted when genotype counts were not presented in the article or could not be calculated, or if an inconsistency was noticed. They were asked to supply raw data (genotype distribution) or clarification. The data points were excluded for quantitative analysis because of insufficient data if no, or only a partial, response was received after two attempts.The credibility of significant associations was assessed according the Venice criteria, based on the published recommendations of the HuGENet Working Groupversus a). In addition, meta‐analysis were performed using dominant (AA + Aa versus aa) and recessive (AA versus Aa + aa) genetic models. Unless stated otherwise, the effect of the minor allele (based on the Ensemble database) was investigatedI2 statistic. Deviation from HWE was calculated systematically, and defined as P < 0·050 using a χ2 test with 1 degree of freedom. Funnel plot analysis and the Harbord statistical testAll analyses were performed using R studio version 3.3.2Fig. Table Table The search resulted in 841 studies potentially eligible for inclusion, of which 96 articles met the inclusion and exclusion criteria  were identified using the allelic contrast genetic model . After credibility assessment, five and three associations were rated as having strong and moderate evidence respectively. The significant associations are discussed in more detail below.Thirty‐one of the 205 extracted variants were reported by at least two articles and were subsequently selected for meta‐analysis. Three data sets were excluded due to insufficient or inconsistent data. A mean of 750 (median 512 i.q.r. 325)) patients and 940 (median 652 (i.q.r. 418)) controls per variant were included in the primary meta‐analyses. Results of the primary and subgroup meta‐analyses, sensitivity analyses and credibility assessments are presented in 25) patieSPINK1 gene in the pancreas function as a trypsin inhibitor, and loss‐of‐function variants are believed to increase autoactivation of trypsinogenSPINK1 gene were described in ten different studies. Only two of the ten studies reported an association of AP susceptibility with the N34S variant . Functionality of the variant remains unclearP = 6·3 × 10−7) (Table P = 1·5 × 10−4), but not of Asians. The association was rated as strong, indicating a highly credible finding.Premature intrapancreatic trypsinogen activation leads to autodigestion and is believed to be a key initiating process in the pathogenesis of pancreatitis) Table . This asP = 3·7 × 10−6) and the Caucasian population (2 studies)  , IL10 and IL18  1β, one of the proinflammatory cytokines that is increased during an episode of APIL1BTable P = 0·018) (Table ) Table . Meta‐anIL6 in vitro, and is associated with susceptibility to juvenile rheumatoid arthritisP = 0·045) (Table P = 0·006) (Table IL‐6 is another proinflammatory cytokine that is correlated with disease severity) Table  and the ) Table . CredibiIL18 have been studied in AP. The −607C/A  and − 137G>C  variants have been shown to increase IL‐18 mRNA levels in expression analyses assaysP = 0·049), but was deemed weak evidence after BFDP correction (Table P = 4·9 × 10−4), and was rated as moderate evidence because of the limited sample size . Owing to heterogeneity, the credibility was rated as weak.C‐C motif chemokine ligand 2 (CCL2), also referred to as monocyte chemoattractant protein 1 (MCP‐1), is involved in the migration of inflammatory cells, and is expressed during pancreatic inflammationIL1B rs1143634), between‐study heterogeneity (IL10 rs1800872), or both (CXCL8 rs4073).Meta‐analyses of other interleukin variants resulted in weak associations, owing to failure to pass correction for multiple testing  was found significantly less often in patients with alcoholic pancreatitis, but not in those with a biliary aetiologyP = 3·3 × 10−7) in allelic contrasts ; however, this was rated as weak evidence owing to the likelihood of being a false‐positive association based on a BFDP value of 0·907 (Table The antioxidant enzyme glutathione S‐transferase (GTS) protects tissues from free radical injury and has four classes: alpha (GTSA), mu (GSTM), pi (GSTP) and theta (GSTT)7 Table .Tables–CXCL8, GSTP1 and TNF (tumour necrosis factor (TNF) α) (Table CXCL8 and GSTP1 variants (both greater than 0·8) indicated a high likelihood of a false‐positive outcome. The BFDP value for the TNF variant was very low, indicating a highly significant association. However, significance was lost when sensitivity analysis excluded one study that deviated from HWE, indicating the presence of genotyping error or population stratificationTLR3, TLR4, TLR6, CD14, NFKBIA, PKA2G7, PPARG and SERPINE1 were found that were not replicated in another study , systemic complications , pancreatic necrosis (HMOX1), mortality (REN) and surgery (TLR2). These are viable candidate genes, and replication is needed . The application of credibility criteria ensures a high likelihood of identifying only true and robust associations.This systematic review and meta‐analysis of genetic association studies in AP found two genetic variants in SPINK1 gene is presumed to be a disease modifier for recurrent and chronic, but not for acute, pancreatitisSPINK1, the relatively large sample size of approximately 1500 patients, and the absence of significant heterogeneity indicates a reliable role in susceptibility to AP and progression to recurrent or chronic disease. Pathogenic variants in the trypsinogen gene PRSS1 are causative for hereditary pancreatitis with a penetrance of 80 per centPRSS1 would also be expected to be associated with AP or recurrent AP. Surprisingly, such an association was not found. However, as PRSS1 variants are associated with early‐onset pancreatitis, they are possibly underrepresented in the present data set owing to the exclusion of studies performed exclusively in paediatric populations. Other possibilities are data paucity, lack of statistical power, and the fact that different variants were studied.The loss‐of‐function N34S variant  in the ALDH2 rs671 variant was found after pooled analyses of two studies by the same authors containing 350 patients and 272 controls of Chinese origin. The functional consequence of this non‐synonymous variant in ALDH2 is a defective enzyme involved in ethanol breakdown, and is common in Asian, but rare in non‐Asian, populations. The A allele is related to alcohol sensitivity, and shown to be protective against alcoholism and alcohol‐associated diseases, such as alcoholic pancreatitisA protective association for the CXCL8 (IL8) and CCL2, but not in other cytokine genes TNF, CXCL8 (IL8) and GSTP1 genes. A previous meta‐analysisCCL2 rs1024611 variant. The present meta‐analysis of the same studies contradicted these results. The present authors explain this discrepancy by an inconsistency in the data of one studyet al.The role of cytokines in the pathogenesis of AP has long been recognizedACE gene was previously reported as having a positive association with AP, but this was not confirmed in the present study; meta‐analysis of the ACE insertion–deletion (rs4646994) in 245 patients and 1455 controls showed a positive association with the insertionA variant in the Fig. No credible associations with disease severity  acute pancreatitis, categorized by gene functionTable S3. Primary meta‐analysis of variants associated with susceptibility for (recurrent) acute pancreatitisTable S4. Subgroup meta‐analysis of variants associated with susceptibility for (recurrent) acute pancreatitis in patients of Caucasian ethnicityTable S5. Subgroup analysis of variants associated with susceptibility for (recurrent) acute pancreatitis in patients of Asian ethnicityTable S6. Sensitivity analysis (excluding studies that deviate from HWE) of variants associated with susceptibility for (recurrent) acute pancreatitisTable S7. Sensitivity analysis (excluding studies that deviate from HWE) of variants associated with susceptibility for (recurrent) acute pancreatitis in patients of Caucasian ethnicityTable S8. Sensitivity analysis (excluding studies that deviate from HWE) of variants associated with susceptibility for (recurrent) acute pancreatitis in patients of Asian ethnicityTable S9. Primary meta‐analysis of variants associated with disease severityTable S10. Subgroup meta‐analysis of variants associated with disease severity in patients of Caucasian ethnicityTable S11. Subgroup analysis of variants associated with disease severity in patients of Asian ethnicityTable S12. Sensitivity analysis (excluding studies that deviate from HWE) of variants associated with disease severityTable S13. Sensitivity analysis (excluding studies that deviate from HWE) of variants associated with disease severity in patients of Caucasian ethnicityTable S14. Sensitivity analysis (excluding studies that deviate from HWE) of variants associated with disease severity in patients of Asian ethnicityClick here for additional data file.


To compare the outcomes of the direct anterior approach (DAA) with the lateral approach (LA) for total hip arthroplasty (THA) patients.Three English databases, PubMed, Embase, and the Cochrane Library, were searched for randomized controlled trials (RCTs) comparing the DAA with LA for THA. Information on the country, sample size, intervention, outcomes, and follow-up were extracted. Meta-analysis was performed using Stata 12.0.P = 0.000) and total blood loss for THA patients . Moreover, the DAA was associated with an increase in walking velocity , stride length , and step length  compared with the LA group. There was no significant difference between groups in the Harris hip score, operation time, transfusion rate, length of hospital stay, and the occurrence of complications.Five RCTs totaling 475 patients  were included in this meta-analysis. Compared with the LA, the DAA was associated with a reduction in the VAS at 6 weeks (weighted mean difference (WMD) = − 0.41, 95% confidence interval (CI) − 0.63 to − 0.19, Current evidence demonstrated a trend showing that the DAA had a better effect on pain relief and blood-saving effects for THA patients. However, considering the number and sample size of the included trials, more large-scale RCTs with high quality are needed to confirm our conclusion. Total hip arthroplasty (THA) is generally considered one of the most successful orthopedic surgical procedures for relieving pain, restoring hip function, and improving quality of life in patients with osteoarthritis (OA) , 2. EnhaThe direct anterior approach (DAA) is an alternative surgical approach for THA. The DAA is less commonly used, although it is gaining popularity in recent years . AdvocatTwo meta-analyses were recently published on this topic. Yue et al.  conducteSeveral more recent RCTs on this subject have been published without conclusive results. Thus, we undertook a further meta-analysis to evaluate whether the DAA is superior to the LA with respect to (1) Harris hip score; (2) pain score at 2 weeks, 6 weeks, and 12 weeks; (3) operation outcomes; and (4) complications and gait characteristics . We hypothesized that the DAA results in lower pain scores and less blood loss without increasing complications.The systematic review and meta-analysis was in accordance with the PRISMA  guidelines  and AMSTThree English databases, PubMed, Embase, and the Cochrane Library, were searched from inception to October 10, 2018, with the following search terms: “direct anterior approach” OR “anterior” OR “direct anterior” AND “lateral approach” OR “lateral” AND “total hip arthroplasty” OR “total hip replacement” OR “THA” OR “THR” OR ““Arthroplasty, Replacement, Hip ”[Mesh]”. There was no language or publication date restriction. We also searched reference lists of included trials and related meta-analyses to identify potentially omitted studies.Participants: patients suffered from osteoarthritis and femur head necrosis and prepared for primary THAInterventions: the intervention group received the DAA for THAComparisons: the control group received LA for THAOutcomes: Harris hip score at final follow-up; VAS at 2 weeks, 6 weeks, and 12 weeks; operation time; total blood loss; transfusion rate; length of hospital stay; complications; and temporal and spatial gait characteristics . Included studies should include at least one of the above outcomes.Study design: RCTs were regarded as eligible for the studyStudy selection was conducted by two reviewers. We removed the duplicates using Endnote X7 software . According to the inclusion and exclusion criteria, we selected the included studies and downloaded the full text for data extraction. Discrepancies were reconciled through discussion or consultation with the author.Data in the included trials were extracted by two independent investigators. Disagreement between the two reviewers was settled by discussion and consultation with a third reviewer. The extracted information included (1) the basic characteristics of the included studies, including the authors, publication year, no. in the DAA and LA groups, mean age, female patients, BMI; (2) outcomes ; and (3) follow-up.Two authors assessed the quality of included studies independently with the risk-of-bias assessment tool outlined in the Cochrane Handbook. Seven domains were evaluated: (1) random sequence generation, (2) allocation concealment, (3) blinding of patients and personal information, (4) blinding of outcome assessment, (5) incomplete outcome data, (6) selective reporting of risk, and (7) other biases.P value < 0.05 was considered statistically significant. The risk ratio (RR) with 95% confidence intervals (CI) for the DAA compared with the LA was calculated for the transfusion rate and the occurrence of complications. Weighted mean differences (WMD) and 95% CI were calculated for continuous variables. A fixed effects model was chosen when there was no statistical evidence of heterogeneity (I2 < 50%), and a random effects model was adopted if significant heterogeneity was found (I2 ≥ 50%). In addition, publication bias was assessed by funnel plots.Stata 12.0  was used for statistical analysis, and a The process of study selection is shown in Fig. Table The risk-of-bias summary and risk-of-bias graph are displayed in Figs. P = 0.689, Fig. I2 = 98.1%); thus, we adopted a random effects model to pool the relevant data. Funnel plot analyses on the Harris hip score demonstrated symmetry, suggesting that bias was minimal  reported Harris hip scores at the final follow-up. There was no significant difference between the DAA and LA groups in terms of the Harris hip score at final follow-up . Compared with the LA, the DAA was associated with an increase in walking velocity  the DAA significantly reduced VAS at 6 weeks and total blood loss; (2) the DAA significantly increased walking velocity, stride length, and step length in analyses of gait characteristics; and (3) the DAA had no benefit for Harris hip score, VAS at 2 weeks and 12 weeks, blood transfusion, length of hospital stay, and complications.Only two meta-analyses of the DAA versus LA for THA have been published. However, differences between ours and previous ones should be noted. First, previous meta-analyses included no more than two RCTs and 106 patients that compared DAA versus LA in THA patients , 12. In Our meta-analysis showed that the DAA has a beneficial role in reducing postoperative pain and blood loss and increasing the function of hip joints. Therefore, the DAA might be the better approach for THA.Furthermore, operator experience and learning curve are equally important factors. All of the included studies have reported that the operators have surpassed the learning curves; thus, the current evidence is limited, and further trials are warranted to identify differences in the learning curve between the two approaches.P > 0.05). However, Mirza et al. [First, we compared the Harris hip scores at the final follow-up. The results showed that the DAA and LA had similar hip function. Zomar et al.  reporteda et al.  revealeda et al.  compareda et al.  found noMoreover, we found that the DAA was associated with a reduction in the total blood loss compared with the LA group. The DAA is an intermuscular and internervous approach without sacrificing muscle injury. Theoretically, the DAA could significantly reduce the approach-related blood loss. Tranexamic acid was routinely administered in THA, and thus, blood loss was not an important factor for hip function. Restrepo et al.  estimateA major strength of our meta-analysis was that we used gait characteristics  to analyze functional outcomes. The results showed that the DAA could increase walking velocity, stride length, and step length. All of the characteristics indicated that the DAA offers significant early advantages in functional recovery compared to the LA.Our meta-analysis also has several limitations: (1) only five studies (475 THAs) were included in our meta-analysis. The statistical efficacy of our results would be more reliable if more studies had been included. (2) Some studies included RCTs that lacked details of the allocation concealment and blinding methods, which may affect the quality of evidence and strength of the recommendations. (3) Follow-ups of these studies were relatively short (maximum follow-up = 2 years), and long-term follow-ups are needed to compare hip function between the DAA and LA. (4) There was substantial heterogeneity between the included outcomes. We performed subgroup analysis and sensitivity analysis to decrease the heterogeneity; however, the overall heterogeneity was not changed after subgroup analysis or after sensitivity analysis.In THA patients, compared with the LA, the DAA was associated with early functional recovery, low blood loss, and lower pain scores. There was no significant difference in Harris hip score or operation time between the DAA and LA groups. Considering the limitations of this meta-analysis, more high-quality RCTs are needed to further identify the effects of the DAA versus LA in THA patients.