I2 = 0%). Compared to normal weight, underweight (BMI < 18.5 kg/m2) and obesity (BMI ≥ 30 kg/m2) were associated with a higher risk of CD, and there was no difference in the risk of U.C. among those with BMI < 18.5 kg/m2 and BMI ≥ 30 kg/m2. There was a significant non-linear association between being underweight and obesity and the risk of development of CD . Conclusions: Obesity increases the risk of IBD development. Underweight and obesity are independently associated with an increased risk of CD, yet there is no evident association between BMI and the risk of U.C. Further studies are needed to clarify the underlying mechanism for these findings, particularly in CD.Background: A growing trove of literature describes the effect of malnutrition and underweight on the incidence of inflammatory bowel disease (IBD). However, evidence regarding the association between underweight or obesity and IBD is limited. The study aimed to assess the association of body mass index (BMI) with a risk of IBD (Crohn’s disease (CD) and ulcerative colitis (U.C.)) incidence. Methods: We systematically searched PubMed/Medline, Cochrane, Web of Science, and Scopus for observational studies assessing the association between BMI and IBD that were published up to 30 June 2020. We estimated pooled hazard ratios (HR) with corresponding 95% confidence intervals (CI). Random effect dose-response meta-analysis was performed using the variance weighted least-squares regression (VWLS) models to identify non-linear associations. Results: A total of ten studies involving 15.6 million individuals and 23,371 cases of IBD were included. Overall, obesity was associated with an increased IBD risk (HR: 1.20, 95% CI: 1.08–1.34,  Globally, there has been a precipitous rise in the incidence of inflammatory bowel disease (IBD) in newly industrialized countries, particularly in younger populations . HistoriToward the end of the twentieth century, obesity emerged as an adverse prognostic factor of various chronic inflammatory diseases, including IBD . Current2 to 0.73% in BMI > 25 kg/m2. Considering the evolving epidemiology of IBD, identifying and controlling the modifiable risk factors offers disease prevention avenues. Thus, this systematic review and meta-analysis aimed to assess body mass index association with the risk of development of IBD (CD and U.C.).The influence of body mass index (BMI) and the development of IBD has previously been documented in epidemiological evidence, with no evidence of the association between increasing B.M.I. and IBD development in the European Prospective Cohort Study (EPIC) . MoreoveWe performed a systematic review and meta-analysis following the guidelines of the Preferred Reporting Items for Systematic Review and Meta-analysis (PRISMA) . A comprAll of the articles and relevant reviews published in English as full peer-reviewed manuscripts were screened for potential missing studies. The process was independently conducted by two reviewers (A.S.B. and J.R.), and the identified studies were evaluated for analysis suitability. After eliminating duplicates, titles and abstracts were screened to exclude the irrelevant studies. All observational cohort studies eligible for the present study were scrutinized thoroughly, and studies that met selection criteria were retrieved for further analysis.Studies were eligible for inclusion according to the following criteria: (1) peer-reviewed articles published in English, with participants aged >18 years; (2) observational cohort studies evaluating the association between BMI and concerning the risk of IBD (CD and/or U.C.); (3) reported the association as odds ratio (OR), risk ratio (R.R.), or hazard ratio (HR), with corresponding 95% confidence intervals (CI) were included.Studies were excluded if they were conducted on children or on patients with a diagnosis of other gastrointestinal disorders, did not incorporate sufficient data of OR/RR/HR to include in the meta-analysis or estimates to compute with 95% CI, and did not stratify according to BMI. Furthermore, in vitro and non-human studies, case reports, case series, editorials, letters, interventional studies, review articles, and non-peer-reviewed articles without sufficient data were also excluded.The information was extracted from each study, independently and blinded by A.S.B., C.C.C., and J.R., and discrepancies were discussed by all authors and resolved through a consensus. The following information was extracted: first author, year of publication, study location, study design, sample size, mean age and gender of the study participants, duration of the follow-up, confounding factors, and summary estimates with the corresponding 95% CI of IBD risk. Data from fully adjusted models were used for the meta-analysis.The quality of each study was assessed according to the Newcastle-Ottawa quality assessment Scale (N.O.S.) , evaluatQ test (Pheterogeneity) and 2I statistics; Pheterogeneity ≤ 0.10 and/or I2 > 50% indicates significant heterogeneity. To examine the potential non-linear association between BMI and IBD risk, we used restricted cubic splines with three knots at the fixed percentiles of 10%, 50%, and 90% of exposure data [p < 0.05. Funnel plots, Begg’s rank correlation test, and Egger’s regression asymmetry test were used to evaluate the publication bias.The meta-analysis was performed using STATA software , using a DerSimonian and Laird multivariate random-effects model to combine the risk estimates of IBD . Normal ure data . StatistThe search strategy yielded a total of 1835 articles, among which 1796 were designated as not pertinent to the study purpose . After eThe core characteristics of the included studies are presented in I2 = 0%) (I2 = 0%).Five studies provided the data of obesity ,11,12,13I2 = 0%)  but not 2 poses a relatively higher risk of CD development   .2 compared with those with normal BMI   .2 [I2 = 0.0)  .2 and suggested the absence of a relationship between obesity and an increased risk of U.C. development , with a significant and high heterogeneity   .1 = −0.0902, p1 = 0.000 and Coef2 = 0.0713, p2 = 0.000)  .p1 = 0.442 and Coef2 = −0.0058, p2 = 0.557)  .All included studies displayed high quality, with N.O.S. scores ranging from 7 to 9. All studies were observed to have drawn the cases and controls from the same community and reported BMI. The majority of the studies provided data from a representative population and detailed the baseline and outcome exposure outcomes. Only one study did not meet the criteria for adequate assessment of the outcome of interest at baseline and did not report the baseline information .Despite the heterogeneity in the outcomes of CD and U.C., the publication bias funnel plots were found to be symmetrical. Additionally, no evidence of publication bias was identified in the primary meta-analysis in Egger’s regression asymmetry test and Begg’s rank correlation test .Worldwide, the ongoing unprecedented shift in the BMI paradigm has led to an inexorable increase in the prevalence of obesity; additionally, approximately 15–40% of IBD patients are obese . On the In the present study, the included articles were contemporary cohorts, all published between 2013 and 2020, which included an equitable number of prospective and retrospective studies, with an appropriately long median follow-up (≈13 years). A total of nine out of ten studies reported the association between BMI and the development of CD ,16,17,202  and BMI ≥ 30 kg/m2 . A clear dose-response association was determined between BMI and CD and U.C. development in the non-linear model. Notably, we observed a paradoxical relationship in which both low and high BMI are strongly associated with a significantly increased risk of CD. Based on our analysis, there is an increased risk of CD in those with low BMI and high B.M.I., compared to normal BMI; additionally, several other studies support these results [Five studies, with a total of 879,959 individuals, were included in the dose-response analysis and reported the risk of CD and U.C. based on the BMI ,12,13,17 results ,13,20.In general, obesity is linked with a chronic, low-grade, pro-inflammatory state, and it is considered an adverse prognostic factor in various chronic inflammatory diseases, including IBD . While tIn this meta-analysis, both underweight and obesity were not associated with the development of U.C. This lack of association between BMI and U.C. may be due to the differences in the U.C. characteristics, such as the ulcerations being limited to the colon, the extent of mucosal damage, differences in the clinical presentation, and the extent of intestinal involvement. However, Kuwahara et al. reported an inverse association between a high BMI and the risk of U.C. exacerbation . AlthougThere are several significant strengths of our meta-analysis that should be noted. Firstly, the present study has provided deeper clarity into the effect of BMI on CD and U.C. by integrating the results from 10 cohort studies, with an unprecedented, extremely large sample size, consisting of 15.6 million participants. This sample size yielded generalizable findings and the opportunity for subsequent dose-response analysis. Second, the current study was free from significant publication bias, low heterogeneity, and symmetrical funnel plots, suggesting that the study findings are reliable. A possible limitation of this study is BMI as a clinical marker; it is indeed worth considering significant confounders that may exist within the compiled studies. Lastly, it is important to note that all the included studies are observational cohorts in design. We must consider the possibility of misclassification bias. Participants in the cohort may suffer an array of additional comorbidities that may have impacted somewhat in the meta-analysis.Underweight and obesity are associated with the development of CD, yet no association between BMI and the risk of U.C. was evident. The dose-response meta-analysis demonstrated that increased BMI was associated with an increased risk of CD, and the curve becomes steeper with increasing BMI. Further studies are needed to discern and clarify the underlying mechanism for these findings.


Autistic spectrum disorder (ASD) is a complex neurodevelopmental disability with a genetic basis, and several studies have suggested a potential role of the reelin gene (RELN) in ASD susceptibility. Accordingly, genetic association studies have explored this potential association, but the results have been controversial thus far. For this reason, we assessed the association of four genetic variants of RELN  with ASD by means of a systematic review and meta-analysis. We retrieved studies comparing the distribution of the above-mentioned genetic variants between ASD patients and healthy controls. A meta-analysis was conducted using a random effects model, and calculations of the odds ratios (ORs) and confidence intervals (CIs) were performed. A sensitivity analysis and tests to determine the heterogeneity of the results were also performed. Eleven previous studies fulfilled the inclusion criteria and analyzed the association of the above-mentioned genetic variants and ASD. We did not find any significant association between the allele or genotype frequencies of the analyzed polymorphisms and ASD, and large heterogeneity was found for the rs736707 polymorphism. Moreover, no significant differences were found between the 5′UTR triplet repeat and this disorder. In light of current evidence, no single genetic variant within this gene is clearly associated with the development of ASD, and ethnic differences may explain part of the observed heterogeneity. Larger studies among different ethnic groups are needed to establish the role of specific genetic variants within RELN in the etiology of this disorder. Autistic spectrum disorder (ASD) is a group of complex neurodevelopment disorders, including autism, pervasive developmental disorder not otherwise specified, Asperger syndrome, and other related conditions ,2. This ASD affects all ethnic and socioeconomic groups. It is often associated with pronounced personal suffering and a heavy burden of care to families and society . Moreove(RELN) maps to 7q22 and encodes a signaling glycoprotein considered to play a key role in the migration of several neuronal cell types and the development of neural connections. Furthermore, Fatemi et al. [RELN in ASD is thus supported by functional studies, as well as genome scans showing ASD linkage peaks in the region that contains this gene.The etiology of ASD is far from completely elucidated, but twin and family studies strongly support a genetic component ,14,15,16i et al.  showed dRELN gene and ASD, but the results have been conflicting. Some studies have found significant associations between ASD and longer triplet repeats in the 5′UTR region of this gene [RELN gene by means of a systematic review and meta-analysis.Several genetic association studies have been performed to analyze the association between genetic variants within the his gene  or certahis gene ,22, othehis gene ,24. GiveRELN . We also analyzed the potential association of ASD with a polymorphic trinucleotide repeat (CGG/GCC) within the 5′UTR region of RELN. The included reports had to include patients with ASD as cases alongside a control group comprised of healthy unrelated subjects. ASD had to be defined by the use of accepted diagnostic criteria ; Diagnostic and Statistical Manual of Mental Disorders, 5th edition (DSM–5); the International Classification of Diseases–10 (ICD–10); the Autism Diagnostic Interview; the Autism Diagnostic Interview-Revised; or the Autism Diagnostic Observation Schedule).In this review, we included case-control studies that analyzed the relationship between ASD and 3 single nucleotide polymorphisms (SNPs) located within Reports published before 3 1July 2020 that fulfilled the inclusion criteria were then identified. For this process, a bibliographic search was undertaken in the following databases: Medline (PubMed), Embase, and Web of Science. The terms used to carry out the search were “autism”, “autistic”, “polymorphism”, “genetic variant”, “reelin”, “reln”, “Polymorphism, Genetic”, and “Autistic Disorder”. There were no language restrictions for this study. The search was complemented by reviewing the references of the included articles. Additional reports were retrieved using the PubMed option "Related Articles" and the Web of Science option “Times cited”.The search and data extraction were independently carried out by three of the authors . Each author imported their search results into a reference manager software (Endnote 6) and removed duplicate references at import. The titles and abstracts were screened independently by the authors  for inclusion and exclusion. Disagreements were resolved by consensus. We recorded the following information: authors’ name(s), year of publication, and country. Moreover, allele and genotype frequencies were extracted or calculated from the raw data.p-values using a random effects model [p-value < 0.05 was considered statistically significant. Cochran’s Q–statistic was used to assess heterogeneity: a significant Q–statistic (p = 0.10) indicated heterogeneity across studies. The I2 statistic was used for estimation of the inconsistency in the meta-analyses (percentage of the observed between-study variability due to heterogeneity rather than chance). The following cut-off points were used: I2 = 0–25%, no heterogeneity; I2 = 25–50%, moderate heterogeneity; I2 = 50–75%, large heterogeneity; and I2 = 75–100%, extreme heterogeneity [Our meta-analysis compared the presence of the above-mentioned allelic variants among patients with ASD as cases versus healthy unrelated controls. The odds ratio (OR) and its 95% confidence interval (CI) were estimated for each paper. The pooled results are reported as the OR with a 95% CI and ts model . A p-valogeneity ,27. In cThis meta-analysis was performed using the computer software packages Review Manager 5.4 and MIX v.1.7. ,29. As iRELN in the cases and controls.The flow of study identification and selection is shown in Moreover, the studies by Krebs et al.  and Li eRegarding the country of origin, three studies were carried out in China ,39,43, tAll studies used genomic DNA extracted from nucleated peripheral blood cells and carried out genotyping using polymerase chain reaction.p = 0.83). Test for heterogeneity: χ2 = 2.85 (p = 0.42); I2 = 0%. B: Distribution of the genotype TT of the exon 50 polymorphism (rs2229864) compared between patients with ASD (cases) and healthy controls under a random effects model. Test for overall effect: Z = 0.49 (p = 0.63). Test for heterogeneity: χ2 = 5.70 (p = 0.13); I2 = 47%. C: Distribution of the genotype TT of the intron 59 polymorphism (rs736707) compared between patients with ASD (cases) and healthy controls under a random effects model. Test for overall effect: Z = 0.12 (p = 0.90). Test for heterogeneity: χ2 = 13.08 (p = 0.02); I2= 62%). The summary and statistics for the association of reelin gene polymorphisms with ASD are shown in p = 0.83; p = 0.63; p = 0.90, respectively). Large heterogeneity was found  when analyzing the rs736707 polymorphism. The comparison of allele frequencies and other models of inheritance also did not yield any significant results.We did not find any significant association between possession of the CC genotype of exon 22 (rs362691), TT of exon 50 (rs2229864), or TT of intron 59 (rs736707) and the presence of ASD (p = 0.26) (p = 0.26). Test for heterogeneity: χ2 = 3.97 (p = 0.14); I2 = 50%). The comparison of specific genotype frequencies also did not show any significant relationship  or the presence of longer triplet repeats in the 5′UTR region of the  studies ,44,45 an studies ,18 stronTo explain our negative results, our meta-analysis indicates that previous studies with different ethnic groups found diverse and even contradictory results when analyzing these allelic variants. For instance, Li et al.  found thRELN [Apart from the risks of false negative results due to ethnicity as a confusion factor and/or low sample sizes, we cannot discount other explanations for our findings. Genetic susceptibility to ASD is probably mediated by many loci, each with a small to moderate effect . TherefoRELN . It has RELN . MoreoveRELN . All theRELN , and we RELN ,46.For our study, we selected the four genetic variants which had been more frequently analyzed in previous studies in order to be able to conduct a meta-analysis ,41,42,43RELN genetic variants and ASD, including data from 1289 patients and 1858 controls. In light of the current evidence, no single genetic variant within this gene is clearly associated with the development of ASD. Larger studies in different ethnic groups are needed to establish the role of specific genetic variants within RELN in the etiology of this disorder.In summary, our meta-analysis represents a comprehensive and up-to-date revision of the association of


Prior studies have examined the influence of MTHFR C677T on autism susceptibility, however, there are no consensus conclusions and specific analyses of a Chinese population. This meta-analysis included a false-positive report probability (FPRP) test to comprehensively evaluate the association of MTHFR C677T polymorphism with autism susceptibility among a Chinese Han population. A large-scale literature retrieval was conducted using various databases including PubMed, Embase, Wan Fang, and the Chinese National Knowledge Infrastructure (CNKI) up to July 31, 2020, with a total of 2,258 cases and 2,073 controls included. The strength of correlation was assessed by odds ratios (ORs) and 95% confidence intervals (95% CIs). MTHFR C677T showed a significant correlation with increased ASD susceptibility under all genetic models . Stratification analysis by region also revealed a consistent association in the Northern Han subgroup, but not in the Southern Han subgroup. Pooled minor allele frequency (MAF) of 30 studies were 45% in Northern Han and 39% in Southern Han. To avoid a possible “false positive report,” we further investigated the significant associations observed in the present meta-analysis using the FPRP test, which consolidated the results. In conclusion, MTHFR C677T polymorphism is associated with the increased risk of autism in China, especially in Northern Han. For those mothers and children who are generally susceptible to autism, prenatal folate and vitamin B12 may reduce the risk that children suffer from autism, especially in Northern Han populations. In the future, more well-designed studies with a larger sample size are expected. Autism spectrum disorder (ASD) involves a constellation of neurodevelopmental disorders featuring impaired repetitive behaviors and deficits in terms of social communication, which are associated with genetic factors and other causes . The preMethylenetetrahydrofolate reductase (MTHFR) gets involved in the process of converting homocysteine into methionine, with the latter one, as a cofactor, playing a critical role in regulating homocysteine concentration in the blood . HomocysPrevious studies have mainly focused on the influence of the MTHFR C677T on autism susceptibility, but the findings are still inconclusive. For example,  reportedThis study undertook a large-scale literature retrieval based on various databases including PubMed, Embase, Wan Fang, and the Chinese National Knowledge Infrastructure (CNKI) up to July 31, 2020. For minor allele frequency (MAF), the search strategy was performed as follows: (“MTHFR” OR “methylenetetrahydrofolate reductase”) AND (“polymorphism” OR “genotype”) AND (“Chinese Han” OR “China Han”). For the association between MTHFR gene C677T polymorphism and the susceptibility to ASD, the predefined search terms are as follows with no language restriction: (“MTHFR” OR “methylenetetrahydrofolate reductase”) AND (“autism” OR “autism spectrum disorders”) AND (“polymorphism” OR “genotype”). The references listed in relevant primary articles were manually checked to avoid missing any other related articles.For MAF, any human studies with a focus on the prevalence of MTHFR C677T polymorphism and ASD in relation to Chinese Han ethnicity were included, regardless of their sample size and report type. To assess the potential genetic association, the inclusion criteria included: (1) case-control studies with the distribution of MTHFR C677T mutation frequencies in autism and non-autism patients; (2) presenting the accurate genotype or allele frequency; (3) standard diagnostic criteria for autism diagnosis; (4) research on Chinese individuals. The exclusion conditions are as follows: (1) not related to MTHFR polymorphism and autism research; (2) repeated publications; (3) previous meta-analysis, case reports, reviews, editorials, and comments; (4) not human beings model studies.Two co-authors (Li and Liu) extracted the data independently from all included studies for analysis, including the first author's surname, publication year, country, region, control source, sample size, genotype frequency of case and control, diagnostic criteria, genotyping method and HWE for controls. When facing discrepancy, we returned to the original studies in discussion with a third reviewer (Che).P < 0.05 indicating disequilibrium. Heterogeneity across the studies was evaluated using both Cochran's Q statistic and the I2 statistic. Specifically, significant heterogeneity was indicated when P was less than 0.10 and I2 was higher than 50% and a random-effects model (the DerSimonian and Laird method) was supposed to be fitted in this case; otherwise, a fixed-effects model using the Mantel-Haenszel was the suitable choice. Furthermore, we conducted sensitivity analysis according to the HWE status of controls, and the subgroup meta-analyses by region were undertaken for particular relationships. We used Begg's funnel plot and Egger's test to detect potential publication bias. The statistical significance of the ORs was determined by Z test, with P < 0.05 indicating a significant difference. Stata 14.0 software was used for the above-mentioned statistical analyses.To measure the strength of correlation between MTHFR C677T polymorphism and autism risk under five genetic models, Odds ratios (ORs), and 95% confidence intervals (95% CIs) were calculated. We also performed a Chi-square test to determine the Hardy-Weinberg equilibrium (HWE) in the control groups, with P < 0.05) between the MTHFR C677T polymorphism and the risk of ASD were “noteworthy,” we further calculated the FPRP value. Although it is suggested to draw on statistical power to detect an OR of 1.5 with an α-level equal to the observed P-value by Wacholder et al.  was shown among all the studies and the pooled MAF estimated by a random-effect model was 42% (95% CI: 37–46%). There was also heterogeneity  among studies on Northern Han , and heterogeneity  was found among Southern Han studies .Five studies , 44, 45 Overall analysis showed that the MTHFR C677T polymorphism increased the risk of autism  under alSensitivity analysis was performed by removing the studies in which the controls were not consistent with HWE and then the pooled OR for the remaining studies was recalculated. The sensitivity analysis suggested similar patterns to the overall analyses , 4.Subgroup analyses showed that when stratified by region, there was a significant association between MTHFR C677T polymorphism and an increased risk of autism in the Northern Han Subgroup under five models .By contrast, a significant correlation was absent in the Southern Han subgroup analyses in each model mentioned above .P < 0.05) detected in the present study were a false positive effect. The results of the FPRP test (see We drew on an FPRP test to investigate whether the significant associations (test see  indicateP < 0.05), and meanwhile, the Begg's funnel plots were consistent with the conclusion , Jacques et al.  can be found in the article/supplementary material.H-FL and C-XL designed the study. C-XL, Y-GL, Y-PC, J-LO, and W-CR performed the literature search, data extraction, and statistical analysis. C-XL drafted the manuscript. Y-GL, Y-LY, and H-FL revised the manuscript. All authors reviewed and approved the final paper for submission and publication.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.


Recently, several studies have found that obesity had a protective effect against varicocele, but no meta-analysis has confirmed this finding. Therefore, we conducted this meta-analysis to investigate the association between body mass index (BMI) and varicocele.We searched for studies in PubMed, Science Direct and the Cochrane Library from inception until February 2018. The association between BMI and varicocele was assessed by pooling the odds ratios (ORs).Eleven eligible studies with a total study population of 1.376.658 participants were included in our analysis. According to BMI, the subjects were defined as belonging to the obese, overweight and underweight groups. Our results showed that the obese group had a lower risk of varicocele when compared with the normal weight group . Additionally, an overweight BMI had a protective effect against varicocele . However, underweight patients had a more than 30% higher risk of varicocele . Furthermore, there was no publication bias in any of the analyses.Our study demonstrates that BMI is negatively associated with the presence of varicocele. Varicocele is present in approximately 15% of the general population. However, more than one-third of men consult doctors regarding infertility, and nearly 80% of secondary infertile men suffer from varicocele . It is w2). It was expected that there would be more than 700 million obese adults and 2.3 billion overweight adults worldwide by 2015 , which is calculated as the weight (kg) divided by the square of the height . Reference lists and conference proceedings were also searched manually to identify possible additional studies.The inclusion criteria were as follows: 1) the topic is varicocele; 2) odds ratios (ORs), relative risks (RRs), hazard ratios (HRs) and standardized incidence ratios with 95% confidence intervals (CIs) were provided or could be calculated; 3) randomized controlled trials or observational studies  published as original studies to evaluate the association between BMI and varicocele; and 4) underweight, obese, overweight or BMI criteria were reported based on the definitions that were established by the Centers for Disease Control. Eligible studies were independently determined by two investigators (Guo Wenbin and Wu Fanglei). Differing decisions were resolved by mutual consensus.Reviews, meeting abstracts, commentaries and editorials were excluded from our analysis. We also excluded the studies if they provided only an estimate of effect, with no means by which to calculate the standard error.A standardized data collection form was used to extract the following information: last name of the first author, year of publication, country of origin, study design, sample size, BMI category, and adjusted effect estimates with 95% CI. Two investigators (Yang Cheng and Huang Zhipeng) independently performed the data extraction.The strength of the relationship between BMI and varicocele was assessed by ORs. ORs were extracted from individual studies and were combined with a fixed-effect model or a random-effect model. Multivariate ORs were used for statistical analysis in preference to the univariate ORs. If the ORs were not directly provided, case and control group numbers were obtained. We first translated the data to ORs for further combination. The ORs from individual studies were transformed to their log [ORs] to stabilize the variance and normalize the distribution before pooling the studies . Pooled For the meta-analysis, both the fixed-effects model (weighted with inverse variance) and the random-effects model were considered based on the level of heterogeneity. Pooled estimates of efficacy were calculated using the Mantel-Haenszel fixed-effects model first . However2 test and I2 statistics, which assessed the appropriateness of pooling the individual study results , ScienceDirect (N=556), and Cochrane Library (N=5). A total of thirty-six potentially eligible studies were identified by the literature search. Three articles were excluded because they were reviews, editorials and responses. We excluded twenty-two studies that did not report the outcome of varicocele or did not provide enough data to calculate the ORs. Finally, we identified eleven full-text articles , 24-31 t2=92.4%). In Rais's study, classification was carried out according to four groups: underweight (<5th percentile); normal weight (5th-84.9th percentile), overweight (85th-94.9th percentile) and obese (≥95th percentile), with normal weight as the reference group. An expanded analysis of the normal weight group included further classification into five percentile groups , with 25-49.9 (the largest group) as the reference group. In the other studies, according to the National Institutes of Health definition, those patients with a BMI of less than 25kg/m2 were categorized as normal weight. Patients with a BMI of 25kg/m2 to less than 30kg/m2 were considered overweight, and those with a BMI greater than 30kg/m2 were categorized as obese.The relationship between overweight and the risk of varicocele was explored in the eleven studies , 24-31. 2=80.3%; Ten , 25-31 s2=81.3%; Four , 28, 31 2=0.0%) and 0.58 , respectively, in the case-control and cross-sectional studies  in the US, 0.65  in Europe and 0.67  in Asia (According to the geographic location analysis, the estimated ORs of varicocele in the overweight group compared with the normal group were 0.79 (95% CI, 0.70-0.90; P<0.001; P for heterogeneity=0.185; I in Asia . Begg's Obesity is associated with significant alterations in the hormonal milieu that can damage the reproductive system , 33. The2 were categorized as normal weight. Patients with a BMI of 25kg/m2 to less than 30kg/m2 were considered overweight, and those with a BMI greater than 30kg/m2 were categorized as obese.Furthermore, the pooled ORs seem to show that the source of the heterogeneity was the study by Rais . When thOur meta-analysis showed an inverse association between BMI and varicocele. Two main theories have been postulated to clarify the inverse relationship between increasing BMI and decreasing occurrence of varicocele. One theory states that varicocele is caused by increased pressure in the left renal vein because it is compressed between the aorta and the superior mesenteric artery . Most reThis is a meta-analysis of observational studies with the limitations inherent in the study design. Therefore, at best, it can demonstrate an association but not a causal relationship. First, most studies calculated the ORs based on data without adjusting for confounding factors. Second, no prospective study could be included in the analysis, which may have biased the results. Third, some of the included studies had different BMI categories, which may confound the pooled results. Furthermore, these studies may have been vulnerable to surveillance bias, as patients with comorbidities would have been more likely to have follow-up imaging studies, leading to the more frequent detection of varicocele than in patients without comorbidities. Future studies that minimize these confounders and biases are needed to confirm this potential causal relationship.Studies have shown that BMI could be a risk factor for left renal vein entrapment. In addition, our meta-analysis showed an inverse association between BMI and varicocele. Thus, for varicocele patients, especially those with lower BMI, attention should be paid to left renal vein entrapment.It is well known that obesity is harmful to human health. The global obesity epidemic parallels a decrease in male fertility. However, the association between BMI and sperm parameters remains controversial. A study found that overweight and obesity are associated with an increased risk of azoospermia and oligozoospermia, which suggests that excess body weight affects sperm production . The invOur study shows a significantly decreased risk of varicocele with increased BMI. However, this potential benefit should not be overemphasized, as obesity itself is harmful to the reproductive system.


ADIPOQ gene in the occurrence of diabetic kidney disease (DKD).  This meta-analysis was performed to obtain a more comprehensive estimation of the role of the single nucleotide polymorphism (SNP) rs2241766 in the I2 statistic, and sensitivity analysis was implemented to test the statistical stability of the overall estimates. Begg's funnel plot and Egger's test were applied to inspect potential publication bias among the included studies.  Relevant studies were identified from digital databases such as Embase, PubMed, Medline, Cochrane Library, Google Scholar, WanFang, and Chinese National Knowledge Infrastructure (CNKI). Odds ratios (ORs) with their corresponding 95% confidence intervals (95% CIs) were pooled by means of fixed- or random-effects models. Interstudy heterogeneity was examined using the Q test and ADIPOQ rs2241766 polymorphism and susceptibility to DKD in the GG vs. TT and GG vs. TT+TG comparisons . After stratification analyses by ethnicity and disease type, a similar trend was also revealed in the Caucasian and African subgroups as well as in the type 2 diabetes mellitus (T2DM) subgroup.  The overall ORs reflected a positive correlation between the ADIPOQ rs2241766 polymorphism may be associated with an increased risk of DKD, especially in Caucasian and African populations as well as in T2DM patients. The  In diabetic patients, microvascular lesions and accelerated atherosclerosis tend to trigger complications leading to severe morbidity . DiabetiADIPOQ), which has been indicated to be linked to susceptibility to cardiovascular disease, metabolic syndrome, and T2DM [ADIPOQ gene mapped to chromosome 3q27 consists of three exons and two introns [ADIPOQ gene has been reported to be a cardiovascular risk factor as well [ADIPOQ gene, and their associations with the risk of DKD in T1DM and T2DM patients have been investigated in diverse populations in many case-control studies, but the results remain inconclusive.One of the candidate genes for DKD is adiponectin , using different combinations of the following keywords: “diabetic kidney disease” or “DKD” or “diabetic nephropathy” or “DN” or “diabetic renal disease” or “DRD” or “diabetic end-stage renal disease” or “diabetic ESRD” or “diabetic renal dysfunction” or “diabetic kidney failure” or “diabetic microalbuminuria” or “diabetic albuminuria” or “diabetic proteinuria” or “diabetic glomerulosclerosis” or “Kimmelstiel-Wilson Syndrome” or “Kimmelstiel-Wilson Disease” or “diabetic complications”, and “adiponectin” or “ADIPOQ rs2241766 polymorphism and susceptibility to DKD; (3) studies providing sufficient information such as genotype frequency for evaluation of odds ratios (ORs) and 95% confidence intervals (95% CIs); and (4) full-text articles of studies with human subjects. Studies meeting any one of the following criteria were considered ineligible for the present meta-analysis: (1) conference abstracts, comments, reviews, case reports, or editorials; (2) insufficient data for OR calculation; (3) no control group; and (4) animal studies.The following inclusion criteria were set for the present meta-analysis: (1) studies including both case and control subjects; (2) studies evaluating the correlation between the We evaluated the quality of all included studies using the Newcastle-Ottawa Scale (NOS). The NOS is composed of 3 aspects: selection, comparability, and exposure, with a total score of 9. According to the final score, the studies could be categorized into high quality (score more than 6), medium quality (score between 4 and 6), and low quality (score less than 4).Data extraction was conducted by two reviewers independently. Conflicting opinions were resolved through discussion to reach a final consensus. The items extracted from each eligible study included the first author's name, publication year, region, ethnicity, disease type, total cases and controls, genotype and/or allele frequencies in case and control groups, genotyping method, and evidence of Hardy-Weinberg equilibrium (HWE) in controls.ADIPOQ rs2241766 polymorphism and DKD susceptibility was determined by calculating pooled ORs and 95% CIs. The interstudy heterogeneity assumption was examined by means of χ2-based Q-statistic and I2 tests. If heterogeneity was significant (p < 0.05 of the Q test or I2 > 50%), the summarized OR estimates were calculated utilizing the random-effects model (DerSimonian and Laird method); otherwise, the fixed-effects model (Mantel-Haenszel method) was applied. Additionally, when the heterogeneity between studies is statistically significant, we would use meta-regression analysis to identify potential sources of such heterogeneity. Begg's funnel plots and Egger's linear regression test were used to evaluate possible publication bias among the included studies. The stability of the combined results was examined by performing a sensitivity analysis, in which each of the included studies was sequentially deleted, and then summary ORs were recalculated to observe alterations between the original and reobtained ORs. The statistical significance of all tests was denoted at p < 0.05.STATA software (version 12.0) was used to conduct all data syntheses in this meta-analysis. The strength of the relationship between the The publication search is described in ADIPOQ rs2241766 polymorphism and DKD susceptibility is illustrated in ADIPOQ rs2241766 polymorphism increased susceptibility to DKD in two genetic comparisons of GG vs. TT  indicated the absence of significant publication bias.Potential publication bias among the included studies was assessed through the visual inspection of Begg's funnel plots accompanied by statistical results from Egger's test. The symmetrical shape of the funnel plots  and p vaADIPOQ gene can prevent vascular remodeling by inhibiting the proliferation and migration of smooth muscle cells and reduce TNF-α production to modulate the inflammatory response of endothelial cells [ADIPOQ gene can affect adiponectin levels, and their contribution to the occurrence of DKD, a common microvascular complication, has been frequently discussed, but conflicting results have been yielded.According to existing evidence, high adiponectin levels in T1DM and T2DM patients may be related to the pathogenesis of diabetic nephropathy –29. The al cells , 31. In al cells . Moreoveal cells . GeneticADIPOQ rs2241766 polymorphism and DKD risk. After data synthesis, we observed a risk-increasing effect of the ADIPOQ rs2241766 polymorphism on DKD susceptibility in the GG vs. TT and GG vs. TT+TG models. The subgroup analyses based on ethnicity and disease type also revealed this positive correlation between the SNP and disease risk in Caucasian, African, and T2DM groups.We therefore performed the present meta-analysis including 3346 cases and 7859 controls to obtain better insight into the relationship between the ADIPOQ rs2241766 polymorphism on susceptibility to DKD remain inconclusive. In a study among patients with T2DM, the GG genotype of the rs2241766 polymorphism was found to be significantly associated with the risk of DKD after adjusting for confounding factors [The results of previous studies regarding the effects of the  factors . In a st factors . Another factors . In cont factors , 17, 23.There are several possible reasons for the above controversy. First, subjects recruited by the above studies belonged to different ethnic groups. Second, confounding exogenous factors such as age, sex, and lifestyle factors were not adjusted in all studies. Third, the limited number of study participants might reduce the authoritativeness of some study results. Of course, some meta-analyses related to our studied topic have already been performed, such as the one by Lin et al.  and one Compared to the above studies, our meta-analysis has many advantages, such as a relatively larger sample size. However, some limitations of the present study should also be mentioned. To begin with, unpublished studies with null results were not included in this meta-analysis, thus possibly introducing certain publication bias, though this bias was not significant. Next, the number of studies for stratification analyses was relatively small, thus affecting the comprehensiveness of the conclusions. Finally, possible interactions of our studied SNP with other relevant factors were not analyzed owing to limited information.ADIPOQ rs2241766 polymorphism might be related to an increased risk of DKD occurrence, which was more evident in Caucasian and African populations as well as among T2DM patients. In view of the abovementioned limitations, these results need to be further verified in the future by studies with larger sample sizes.Overall, the present meta-analysis indicated that the


Metabolically healthy obesity (MHO) is defined as obesity with less than two parameters of metabolic abnormalities. Some studies report that MHO individuals show similar risk of cardiovascular disease (CVD) compared with metabolically healthy non-obese (MHNO) individuals, but the results are conflicting. Coronary artery calcium (CAC) reflects the extent of coronary atherosclerosis and is a useful tool to predict future risk of CVD. The objective of this meta-analysis was to investigate whether MHO is associated with elevated risk of CAC.We searched Cochrane, PubMed, and Embase up to April 19, 2019. Prospective cohort and cross-sectional studies examining the association between MHO subjects and CAC were included with MHNO as the reference. Pooled odds ratio (OR) and 95% confidence interval (CI) were calculated using random-effect models. Subgroup analysis and meta-regression were applied to define possible sources of heterogeneity. We conducted this research following a pre-established protocol registered on PROSPERO (CRD 42019135006).I2 = 39%). In the subgroup analysis, the risk associated with MHO participants was significant in cohort studies , and borderline significant in cross-sectional studies. The risk of CAC was also significant in MHO participants defined by Adult Treatment Panel III (ATP III) . The univariate meta-regression model showed that age and smoking status were possible effect modifiers for MHO and CAC risk.A total of nine studies were included in this review and six studies with 23,543 participants were eligible for the meta-analysis. Compared with MHNO subjects, MHO had a higher odds of CAC (OR 1.36, 95% CI [1.11 to 1.66]; Our meta-analysis showed that MHO phenotypes were associated with elevated risk of CAC compared with MHNO, which reflects the extent of coronary atherosclerosis. People with obesity should strive to achieve normal weight even when only one metabolic abnormality is present. Worldwide obesity rates have nearly tripled in the last 40 years, and to date more than one third of the global population is overweight or obese . ObesityHowever, not all obese individuals have metabolic abnormalities. This group includes those without insulin resistance, dyslipidemia, glucose intolerance, hypertension, or high inflammatory status . Recent Coronary artery calcium (CAC), detected by coronary multidetector computed tomography (MDCT), reflects the extent of subclinical atherosclerosis and suggests the presence of CVD . PreviouThis systematic review and meta-analysis were conducted following a pre-established protocol registered on PROSPERO (CRD 42019135006) and reported in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines  Table S. All autWe performed a systematic literature search using the Medline, EMBASE, and Cochrane library databases supplemented with the manual review of the reference list of obtained articles up to April 19, 2019. We used different combinations of the following Medical Subject Headings (MeSH) terms:  and . Full search strategies are shown in We included all eligible publications that satified our inclusion criteria: (1) published observational cohort and cross-sectional studies investigating MHO and CAC conducted on adults; (2) classification of obesity and non-obesity by waist circumference (WC) in addition to the body mass index (BMI) shown in the pre-established protocol; (3) reporting of criteria used to define metabolically healthy/unhealthy (MH/MUH) phenotypes; (4) presence of a reference group with metabolically healthy non-obese or normal weight individuals; (5) reporting of the presence, extent, new development, or progression of CAC, assessed by electron beam computed tomography (EBCT) or MDCT. Studies were excluded if they (1) were duplicate publications, (2) topically irrelevant, (3) did not compare MHO and CAC with MHNO or metabolically healthy normal weight (MHNW) people, or (4) were literature reviews, republished data, case reports, dissertations, or conference abstracts.We extracted the following information: first author’s name, year of publication, description of the study population , study design, definition of metabolically healthy and obese, diagnostic methods of CAC, adjusted variables, and main outcome presented with OR (95% CI) using MHNO as the reference.The quality of the included studies was assessed independently by two authors, Chien-Yu Lin and Szu-Ying Tsai, using the Newcastle Ottawa Scale (NOS). The quality assessment tool rates each study in three domains—selection, comparability, and outcome—using a star system ranging from zero to nine stars in cohort studies and from zero to ten stars in cross-sectional studies . If two Q test and I2 statistics  with 95% CIs for estimating the risk of CAC progression in MHO compared with MHNO. We used statistical computing software R, Version 1.1.456, primarily using the Comprehensive R Archive Network (CRAN) package “metagen”for our meta-analysis . We emplatistics  and furtatistics . With reevidence . We perfue < 0.1 .The flowchart of article selection is shown in The general demographic characteristics of subjects in the nine included studies in the systematic review are summarized in Obesity was defined by a BMI ≥ 25 in the Korean studies  and ≥30 For evaluating outcomes, the three cohort studies used CAC progression as expressed by a categorical change or change in the absolute coronary artery calcification score (CACS) . The othAll three cohort studies achieved at least eight out of nine stars on the NOS quality assessment scale shown in . The croI2 = 39%, I2 = 0%, forest plot shown in I2 = 69%, forest plot shown in I2 = 0%, forest plot shown in I2 = 68%, forest plot shown in For CAC risk analysis, six observational studies with 23,543 subjects were pooled for the meta-analysis . Particip = 0.31) and was borderline significant with age and smoking status . We also excluded one study at a time and re-calculated the overall effect estimate, and the results were not changed. No significant publication bias was detected by Egger’s tests (p = 0.47) and there was no substantial asymmetry in the funnel plot  of 2.00 (95% CI [1.79–2.24]) compared with healthy normal-weight individuals. The results appeared much stronger during the long-term follow-up period of more than 15 years . AnotherIn the subgroup analysis of our study, MHO participants in cohort studies showed a higher odds of CAC progression, and it is likely that MHO is associated with CAC prevalence in the cross-sectional studies. Of the studies included in our meta-analysis, three were cross-sectional and only offered a snapshot of subclinical coronary atherosclerosis in MHO . Althoug2); and (f) high-sensitivity C-reactive protein (hs-CRP) >90th percentile (>0.1 mg/L). Wildman criteria also included HOMA-IR, the core concept of metabolic syndrome, and CRP, the best biomarker of vascular inflammation and predictor of CVD events (I2 = 0%), the definiton of CAC progression differed in the three included cohort studies. CAC progression was defined by absolute CAC change in two cohort studies  from the criteria in our study. However, a previous meta-analysis showed that compared with MHNW, MHO participants not expressing any of these metabolic factors showed no significantly increased CV risk . In one Age, and smoking status were possible effect modifiers in our study. Previous studies indicated that CAC risk increased with age, whereby the extent of CAC was greater in men compared with women up to the age of 60 . In contβ-cells. These locations are associated with the risk of developing CVD , an indicator of arterial wall thickness and atherosclerotic disease progression. Bobbioni-Harsch et al. performed a three-year progression evaluation showing significantly elevated CCA-IMT in MHO compared to normal body weight  which isOur study showed a significant association between MHO and elevated risk of CAC, which in turn reflects the extent of coronary atherosclerosis. People with obesity should strive to achieve normal weight even when only one metabolic abnormality is present.10.7717/peerj.8815/supp-1Data S1Click here for additional data file.10.7717/peerj.8815/supp-2Table S1From: Moher D, Liberati A, Tetzlaff J, Altman DG, The PRISMA Group (2009). Preferred Reporting Items for Systematic Reviews and Meta-Analyses: The PRISMA Statement. PLoS Med 6(7): e1000097. doi:10.1371/journal.pmed1000097 For more information, visit: www.prisma-statement.org .Click here for additional data file.10.7717/peerj.8815/supp-3Table S2Click here for additional data file.10.7717/peerj.8815/supp-4Table S3-1Each star represents if individual criterion within the subsection was fulfilled.Click here for additional data file.10.7717/peerj.8815/supp-5Table S3-2Each star represents if individual criterion within the subsection was fulfilled.Click here for additional data file.10.7717/peerj.8815/supp-6Figure S1p = 0.08)Each bubble represented a study and bubble size represented the sample size of the study. The regression line showed a non-significant trend of increasing risk with age increased. Each bubble represented a study and bubble size represented the sample size of the study. The regression line showed a non-significant trend of declining risk with women proportion increase. Each bubble represented a study and bubble size represented the sample size of the study. The regression line showed borderline significant trend of decreased risk with smoke proportion increased. (Click here for additional data file.10.7717/peerj.8815/supp-9Figure S4Click here for additional data file.10.7717/peerj.8815/supp-10Supplemental Information 1Click here for additional data file.


This study is aimed at investigating the overall association between the LTA A252G polymorphism and the risk of sepsis/septic shock and sepsis-related mortality.  The association between the lymphotoxin-LTA A252G polymorphism and risks of sepsis, septic shock, and mortality. The relevant data were extracted, and statistical analyses were performed using the Revman 5.0 and STATA 12 software.  We searched the PubMed and EMBASE databases to identify studies that investigated the association between the LTA A252G polymorphism showed no significant association with sepsis risk  or with sepsis shock risk . However, in the subgroup analyzed by ethnicity, the LTA A252G polymorphism significantly decreased sepsis risk in the Asian population for the recessive model  but not in the Caucasian population. Moreover, comparisons between sepsis patients who survived and those who did not suggested that the LTA A252G polymorphism decreases the risk of mortality .  A total of 32 publications were included in the meta-analysis. The results demonstrated that the LTA gene decreased the risk of sepsis in Asians and may reduce mortality in septic individuals. Our results suggested that the A252G polymorphism in the  Despite advanced treatments, the sepsis mortality rate, which is around 20-30%, remains hard to ignore . AlthougLTA+252 A allele frequency was the most predominant allele in most of the world populations; and the LTA+252 G allele was associated with the outcome of different diseases [LTA A252G polymorphism  was associated with inflammatory response, including sepsis. The LTA A252G polymorphism has been reported as a sepsis susceptibility variant; however, the results have been inconclusive. In 2011, a meta-analysis was performed to assess the overall association between sepsis risk and the LTA A252G polymorphism [LTA A252G polymorphism on sepsis risk, and these studies should also be included. To obtain a more reliable and precise conclusion about the association between the LTA A252G polymorphism and sepsis/septic shock risk and sepsis-related mortality, we performed this updated meta-analysis with accurate data and current eligible studies.Genotype frequency showed that the diseases . The higdiseases . LTA exediseases . It has LTA A252G polymorphism and sepsis/septic shock risk and mortality, updated on July 14, 2020. The search terms used were as follows: “sepsis or severe sepsis or septic shock” in combination with “polymorphism or variant or mutation” and “lymphotoxin-α or LTA or tumor necrosis factor-β or TNF-β.” The inclusion criteria were as follows: (1) they were case-control genetic studies, (2) they evaluated the association between the LTA A252G polymorphism and sepsis/septic shock risk or mortality, and (3) the genotype distributions for cases and controls were sufficient to estimate the odd's ratio (OR) with a 95% confidence interval (95% CI). The exclusion criteria were as follows: (1) abstracts, letters, and review articles; (2) genotype frequency not shown, and (3) repeated or overlapping data.We carried out a literature search in the PubMed and EMBASE databases to identify studies that investigated the association between the Two independent authors checked all the included studies and reached a consensus on every item. The following data were extracted from the included studies: author, year of publication, country of origin, ethnicity, sepsis source, sepsis definition, gene assay method, total number and distribution of genotypes, and genotyping methods.χ2 test. A p value of <0.05 indicated deviation from HWE. The strength of the association between the LTA A252G polymorphism and sepsis risk was assessed by the odds ratio (OR) with its corresponding 95% confidence interval (95% CI). We applied a random effects (DerSimonian and Laird method) or fixed effects model (Mantel-Haenszel method) to pool the OR values according to the results of the heterogeneity examination. Heterogeneity was assessed by a χ2-based Q statistic and I2, and a p value of <0.10 was statistically significant. For p < 0.10, the pooled OR was calculated using a random effects model. Otherwise, a fixed effects model was used. The I2 statistic was used to estimate the degree of heterogeneity, and a value > 50% was considered an indication of a large degree of heterogeneity. The significance of the pooled OR was evaluated by a Z-test, and a p < 0.05 was statistically significant. The dominant genetic model (GG+GA vs. AA), recessive model (GG vs. GA+AA), codominant model (GG vs. AA), heterozygote model (GA vs. AA), and allele model (G vs. A) were used to pool ORs and assess the association of each genotype with the risk of sepsis. Subgroup analyses were performed for accordance with HWE, ethnic group, septic shock, and mortality of sepsis population.Hardy-Weinberg equilibrium (HWE) was tested using Pearson's Publication bias was assessed by Begg's funnel plots and Egger's test. Sensitivity analyses indicating the reliability of a meta-analysis were conducted to identify the potential influence of individual data sets to the pooled OR. All statistical analyses were performed using the Revman 5.0 software  and the STATA 12.0 software .LTA A252G polymorphism and sepsis risk was reported in 25 studies [A total of 225 studies were identified after an initial search of the PubMed and EMBASE databases. After reading the full-text, one article  that was studies , 19–42,  studies , 38, 40, studies , 38–41. +GA vs. AA for all 24 studies and the χ2 value was 59.5 with 23 degrees of freedom (p = 0.27). In addition, the I-square value, another index of heterogeneity, was 61%. A fixed effects model was used to pool the data. The overall OR was 0.92 (95%CI = 0.79–1.07), and the overall effect Z value was 1.10 (p = 0.27) for the GG+GA vs. AA model (vs. GA+AA), codominant model (GG vs. AA), and allele model (G vs. A), which did not indicate any associations with the risk of sepsis, are listed in For an overall analysis of sepsis risk, we analyzed the heterogeneity of GGAA model . The resLTA A252G polymorphism and sepsis risk . In the subgroup analysis by ethnicity (Caucasian and Asian), no association was identified between the LTA A252G polymorphism and sepsis risk in Caucasians  and Asians . However, the recessive model for the Asian populations showed decreased risk of sepsis   .LTA A252G polymorphism on septic shock risk [p = 0.91 for GG+GA vs. AA). Furthermore, a total of 21 studies had determined the association between the LTA A252G polymorphism and the mortality of sepsis [LTA A252G polymorphism significantly decreased the mortality risk of sepsis patients . After sequentially excluding each one of the 25 studies that assessed the overall relationship between the LTA A252G polymorphism and sepsis risk, statistically similar results were obtained, suggesting the results of this meta-analysis were stable .A sensitivity analysis was performed to evaluate the stability of the individual data to the pooled OR  could decrease the mortality rate in septic patients.Since sepsis, in severe cases, can progress to septic shock and death, we also analyzed the association between the LTA A252G polymorphism and sepsis risk, indicating that this polymorphism might not have been included in those GWAS arrays. Thus, future studies are needed to further assess and validate our results.GWAS is the most appropriate method to identify susceptible genes for sepsis , and manHeterogeneity and publication bias play a determining role in the reliability of the results in a meta-analysis. Significant heterogeneity was detected in some comparisons; however, this may be due to study design differences among the included studies. When significant heterogeneity was found, a random effects model was applied for analysis. In addition, the genetic distribution of GG, GA, and AA could not be extracted independently in some cases, probably partly contributing to the existence of heterogeneity.Publication bias and sensitivity analysis constitute an essential index for the quality and reliability of the study. Publication bias was analyzed using Begg's funnel plots and Egger's test in our study. The results indicated the reliability of our meta-analysis.LTA A252G polymorphism with sepsis risk. However, this study had some limitations. First, since our literature search was conducted only in the selected databases, we might have missed relevant studies deposited in other databases. Second, since we only included published studies written in English, studies in other languages were excluded. Third, most of the included studies were conducted in Caucasian and Asian populations; therefore, the results may only be applicable to these populations. Hence, future studies are warranted to explore these associations further, particularly in African-American, African, and Latin populations. Nevertheless, this meta-analysis has made an important contribution to this field. A comprehensive evaluation of the association between the LTA A252G polymorphism and sepsis risk is more powerful than a single study. Furthermore, the reliability of this meta-analysis was confirmed by heterogeneity, publication bias, and sensitivity analyses.Hitherto, this is the most specific and comprehensive meta-analysis to investigate the association of the LTA gene and sepsis risk. Our results suggested that the LTA A252G polymorphism was significantly associated with a decreased risk of sepsis in Asian populations and with a decreased risk for mortality among septic individuals.To our knowledge, this is the most comprehensive meta-analysis to assess the relationship between the A252G polymorphism in the


GSTP1, GSTM1 and GSTT1 gene polymorphisms in susceptibility to Chronic Myeloid Leukaemia (CML). The development of cancer results from an imbalance between exposure to carcinogens and the capacity of various enzyme systems engaged in activation or in the detoxification of xenobiotics. The aim of the present study is to investigate the association of GSTM1 and GSTT1 analysis with PCR and GSTP1 analysis with PCR-RFLP. A total of 200 CML patients and 100 controls were enrolled in a case-control study with GSTT1 null genotype was significantly higher among CML patients suggesting that this genotype is associated with an increased risk of CML. It was found in 42% of cases as compared with 21% of the controls, . The presence of the GSTT1 genotype may thus be considered a protective factor for CML. The frequency of individuals carrying GSTM1 null genotype was slightly higher in the control group but this difference was not statistically significant. The GSTM1 null genotype was present in 35% of control cases and 34% of the CML patients, . Individuals with a combined GSTM1 null/GSTT1null genotype had an estimated 2.85-fold increased risk of CML, but no associated risk between GSTP1 Ile 105 Val polymorphism and CML was found . The GSTP1 and GSTM1 with susceptibility to CML was found. GSTT1 genotype may be a protective factor for CML, while the null genotype shows association with developing CML.No association between  Chronic Myeloid Leukaemia (CML) is a myeloproliferative disorder but a definite mechanism leading to this carcinogenesis is not yet completely understood . It is known that environmental exposure to cytotoxic and genotoxic agents derived from benzene may be associated with increased risk of CML . And genetic susceptibility studies of CML may serve to identify populations at risk here, together with its important disease mechanisms .The development of cancer results from an imbalance between exposure to carcinogens (endogenous and exogenous) and the capacity of various enzyme systems engaged in the activation or detoxification of Xenobiotics. Inter-individual genetic variation in xenobiotic metabolizing enzymes has been associated with cancer development,  and since such a metabolism constitutes an important line of defense against a variety of carcinogens, inherited differences in the capacity of these enzymes may be an important genetic factor in cancer susceptibility . GSTP1, GSTM1, and GST1 . Polymorphisms in these genes lead to the absence or decreased detoxification ability of enzymes, their dysfunction, and finally may impact on the risk of cancer development and heterogeneous drug responsiveness . Xenobiotic metabolizing enzymes (XMEs) constitute one of the first lines of defence and they play a central role in the metabolism, elimination, and detoxification of xenobiotics or exogenous compounds introduced to the body . Enzymes within the multiple enzyme system are classified in two categories namely Phase I and Phase II. The latter constitutes the theme of this study and its major detoxifying enzymes are Glutathione S-transferases (GSTs) which enable a wide variety of functions. GSTs fall into two distinct super-families: membrane bound microsomal GSTs and the soluble or cytosolic GSTs. Some genes of cytosolic enzymes play a crucial role in the detoxification of activated carcinogens and implications in cancer progress, particularly GSTP1 gene possesses two variations in coding region, an A-to-G transition at105 codon and a C-to-T transition at 114 codon . The GSTP1 Ile105Val polymorphism at nucleotide 313 in exon 5 of GSTP1 lead to an amino acid substitution of isoleucine (Ile) by valine  at amino acid position 105  . This substitution potentially diminished the ability to detoxify certain carcinogens which would induce DNA adducts, ultimately leads to carcinogenesis . The frequency of the homozygous mutant genotype  ranged from 4% to 16% among White, 4% to 5% among Asian, and 19% for African-American ethnicities .GSTM1 polymorphism has been identified with three alleles; GSTM1*0, GSTM1*A and GSTM1*B . The former is a null allele consisting of the complete deletion of the GSTM1 gene. Individuals who are homozygous for this allele are unable to produce the GSTM1 protein. The null variant of GSTΜ1 is of particular interest as a plethora of studies have demonstrated the difference in its susceptibility, exposure to environmental toxicants, and resistance to chemotherapy treatment . The frequency of the GSTM1 null genotype ranges from 23% to 48% in African populations, 33% to 63% in Asian populations, 39% to 62% in European populations, and 23% to 62% in U.S. populations . GSTT1 gene is polymorphic and has two alleles, the GSTT1*1 and the GSTT1*0 . The former is an active allele, having a crucial role in the Phase-II biotransformation of a number of drugs and industrial chemicals. The latter is a non-functional allele arising from the deletion of the GSTT1 gene, with null GSTT1 (0/0) phenotype individuals being unable to form the GSTT1 protein . The homozygous GSTT1 null phenotype has been described in different populations and shows wide variation . The frequency of the GSTT1 null genotype has been described in different ethnic groups with varying degrees as follows: 22% to 29% in African-Americans, 16% to 64% in Asian individuals, 15% to 27% in whites, 10% to 21% in European populations and 10% to 12% in Mexican-Americans . Therefore these alleles and the previous ones have taken a place in this study. The GSTP1 Ile105Val, GSTM1 and GSTT1 polymorphisms in the susceptibility to CML was investigated by different studies, but with conflicting results . Therefore, this study aimed to determine the frequency of genetic polymorphism in GST  genes and to ascertain their association with CML within the Sudanese population.The association between This cross-sectional, case-control study was conducted at the Radiation and Isotopes Center of Khartoum, (RICK). The study included 200 patients diagnosed with CML, 68 (34%) females and 132 (66%) males plus 100 age matched healthy controls, 49 (49%) females and 51 (51%) males; 300 participants in total. All CML patients were in chronic phase with exception of one patient who was in an accelerated phase.The Institutional Ethical Committee of Al-Neelain University approved the study and written consent was obtained from all the participants involved. Patients suffering from any other disease such as chronic myelomonocytic leukaemia and other myeloproliferative disorders were excluded. The control group consisted of healthy unrelated volunteers without a medical history of cancer and all patients and controls were of Sudanese ethnicity.DNA extraction: Venous blood samples (3ml) were collected into EDTA containing vacutainers. Genomic DNA used for polymorphic analysis was extracted by guanidine chloride method and isolated DNA was stored in Tris EDTA buffer at -20°C till use.GST P1 polymorphism: The genetic polymorphism analysis for GSTP1was determined by using the polymerase chain reaction/ restriction fragment length polymorphism (PCR/ RFLP) method. Primer pairs used were synthesized from Macrogen) F5′-ACCCCAGGGCTCTATGGGAA-3′ and R5′TGAGGGCACAAGAAGCCCCT-3′). A three-step PCR was standardized using Sensoquest thermocycler, involving initial denaturation at 95°C for 3 min, followed by 35 cycles of denaturation at 94°C for 30 sec, annealing at 61°C for 30 sec, and extension at 72°C for 40sec – with a final extension at 72°C for 5 min.Amplification products corresponding to 176bp were then visualized after electrophoresis in an ethidium-bromide-stained 2% agarose gel and then subjected to restriction enzyme analysis with BsmA1 (New England BioLabsInc). Quantities 0.2 μl of BsmA1enzyme, 2 μl of enzyme buffer, 5 μl of the PCR product and 2.8 μl of double distilled water were then incubated overnight at 37°C. Three banding patterns were observed on 3% agarose gel stained with ethidium-bromide and these were; 176 bp bands corresponding to the AA , 176, 91 and 85 bp bands corresponded to the AG  and 91, 85 bp bands that corresponded to the GG .GSTM1 and GSTT1 polymorphism: The polymorphic deletion of GSTM1 and GSTT1 genes were genotyped using the multiplex PCR approach. Primers used for GSTM1 and GSTT1 amplification were synthesized from Macrogen: F5’-GAA CTC CCT GAA AAG CTA AAG C-3’; R5’-GTT GGG CTC AAA TAT ACG GTG G-3’ and F5’-TTC CTT ACT GGT CCT CAC ATC TC-3’, R5’-TCA CCG GAT CAT GGC CAG CA-3’, respectively. An 268-bp fragment of β-globin gene amplified by F5’-CAA CTT CAT CCA CGT TCA CC-3’, R5’- GAA GAG CCA AGG ACA GGT AC-3’ primers was used as internal positive control. GSTM1 and 480 bp for GSTT1 PCR products. PCR products for the genotyping of polymorphisms were visualized by 2% agarose gel electrophoresis with ethidium bromide. The absence of β-globin amplification here indicated a failure of PCR reaction. The thermocycling procedure , involved initial denaturation at 94°C for 4 minutes, followed by 35 cycles of 1 minute at 94 °C, 45 seconds at 55°C, 1 minute at 72°C and a final extension for 10 minutes at 72°C. Genotyping of the genes (null genotypes) is revealed by the absence of the 219 bp for Statistical analysis: Statistical analysis included descriptive statistics of the mean using standard deviation. The Odds ratio (OR) with a confidence interval (CI) of 95% was calculated by logistic regression. Pearson’s chi-square test was then used to compare genotype distribution between patients and control, with p-values less than 0.05 being considered as statistically significant. GSTP1, GSTM1 and GSTT1 genotypes in CML patients and controls are shown in The distribution of the GSTP1 Ile105Val, heterozygous  and the wild genotype of GSTP1 (Ile/Ile) forms were 2%, 36% and 62%, respectively. The homozygous mutant type  showed no significant difference between patients and controls .In the Control, the homozygous  of The frequency of individuals carrying the GSTM1 in patients and controls were 44% and 21.7% respectively. GSTM1 null genotype frequency was found to be slightly higher in the control group, (35% as opposed to 34% in CML patients), but this difference was not considered to be statistically significant .The GSTT1 was found in 57.5% of CML patients and 79% of the Control but the frequency of individuals carrying the GSTT1 null genotype was significantly higher among CML patients, 42% compared to 21% of the Control; . The GSTT1 genotype presence may thus be considered a protective factor for CML.GSTM1 and GSTT1 genotypes in CML risk were also conducted. Individuals with a combined GSTM1 null/GSTT1 null genotype had an estimated 2.847-fold increased risk of CML over individuals with a GSTM1 present /GSTT1 present genotype . In contrast, the GSTM1 present /GSTT1 null and GSTM1 null/GSTT1 present genotypes were not associated with a CML risk , see The combined effects of GSTP1 Ile105Val polymorphism. No significant difference was found between patients and the control and the study also revealed no association between GSTP1 Ile 105 Val polymorphism and the risk of developing CML:. Several previous studies have reported an association between the polymorphisms of the GSTs genes and the susceptibility of developing certain types of cancer. Such cancers include lung cancer , ovarian cancer  and breast cancer . Previous studies have reported about GSTs association with heamatological malignancy, but with conflicting findings as mentioned previously . In the present study, 3.5% of CML patients and 2% of the control had the homozygous type  of GSTP1Ile105Val and CML susceptibility  . Weich and co-workers also reported no meaningful association between GSTP1Ile105Val and CML risk . Several studies disagree with our findings in stating that there may be a relationship between GSTP1 Ile 105 Val polymorphism and CML development/prognosis . Some of these studies relied on smaller sample size and others on very broad confidence levels. Such contrary findings may also not have taken into account the geographical and ethnic differences in the type of environmental carcinogens samples the patients may have been exposed to .This finding was in agreement with a study conducted in Turkey by Karkucak et al who found no evidence between GSTM1 and GSTT1 lack the activity of respective enzymes. This makes the reactivity of electrophilic substrates low and therefore affects the function within cellular macromolecules. . The GSTM1 null and GSTT1 null appear to be associated with a significant risk of several types of cancers, such as hematological neoplasm  and breast cancer . It has been shown that the risk of ALL is doubled in patients who carry the GSTM1 deletion . The GSTT1 null genotype was also demonstrated as a risk factor for both AML and ALL in some ethnic groups .Homozygotes for the null allele (deletion) of GSTM1 null genotype and the risk of CML was found here: . However, Muddathir and his co-workers found that GSTM1 null genotype was associated with increased risk of CML among Sudanese. This is not unexpected because Sudan is multiethnic with an admixture of Arab and African lineages which may be associated with differences in genetic makeup. The tribal differences within the participant populations in these two studies, irrespective of different in sample sizes, may contribute to this variation in the results. In contrast, Taspinar et al and Hishida et al claimed that GSTM1 polymorphism was not associated with the risk of CML, which was in agreement with our study . Al-Achkar and co-workers however, concluded that GSTM1 null genotype was associated with increase risk of CML:  . An interesting result was observed by Lordelo et al who found a positive association between CML risk with GSTM1 present genotype and that the GSTM1 null genotype decreased this risk . Interestingly, no association between GSTT1 null genotype in CML patients was significantly higher than in the control . GSTT1 null genotype frequency posed a 2.781 fold increase in the risk of CML, compared to those possessing both alleles. Therefore, the GSTT1 genotype may be a protective factor for CML, whilst the null genotype showed an association with the development of CML. The above findings were in agreement with a recent study conducted in Sudan . Many studies have focused on the relationship between GSTT1 polymorphism and the risk of CML in diverse ethnic groups . These findings were in agreement with this study. In contrast however, a Japanese study reported no significant association between GSTT1 and the risk of CML  . In the present study, we observed that the frequency of GSTT1 is involved in the metabolism of ethylene oxide (EO) - a genotoxic agent capable of producing heritable translocations and increasing the frequency of spontaneous chromosome abnormality. Presuming that EO is higher in GSTT1 null individuals, it was hypothesized that the loss of the GSTT1 gene may be an influencing factor in the production of the Philadelphia chromosome (Ph) associated with CML . GSTM1 null and GSTT1 null genotypes, and thus, individuals carrying the null genotype of both are at a higher risk to CML: . It might be inferred from the data that the both genes (GSTM1 and GSTT1), act in a synergistic way and are important to the detoxification system . When these genes lack their enzyme activity and become inactive there is an increased opportunity for DNA damage, resulting in the risk elevation of the double null genotype to CML , as found in this study. The risk elevation of this double null genotypes was also in accordance with the Al-Achkar and co-worker study . Taspinar et al., (2008) also suggested that the association between the GSTT1 or GSTM1 genotype and CML depends on relative expression levels. Therefore the current result suggested a distinguished haplotype which was found with a higher susceptibility in having CML as detected in not many ethnic groups.A combined analysis was conducted to assess the role of polymorphic variants on CML risk. We observed a significant interaction between the GSTP1 mutant allele Val in the advanced disease state (accelerated and blast crises), as compared to the chronic phase . Furthermore, Lourenco et al reported the frequency of the GSTM1 null genotype being lower in patients in the accelerated phase or with blast crisis than those patients in the chronic phase: . In contrary, Bənescu et al., (2014) found no association between GSTM1 null or GSTT1 null and the clinical phases of CML. In the present study, it was not possible to compare GSTs polymorphism within the clinical phases of CML, because only one sample was found to be on accelerating phase with no blast crisis. However, A study has reported significant increases in the frequency of the GSTT1 null genotype in CML patients was significantly higher than in the control and thus the GSTT1 genotype may be assumed a protective factor for CML, with the null genotype associated with the development of this disease. Additionally, no association between GSTP1 and GSTM1 and the susceptibility to CML was found. This study may provide a basis for further more extensive testing in a larger Sudanese population.In conclusion, this study found that the percentage of the


Background: Snoring is a common condition. Previous studies have reported the relationships between snoring and metabolic syndrome (MetS) and/or its five components: hypertension, hyperglycemia, low-high density lipoprotein (low-HDL), high-triglyceride level, and abdominal obesity. However, conclusions have been inconsistent, and there has been no comprehensive summary on this. Therefore, we performed a systematic review on the relationships between snoring and MetS, including each of MetS' components.Methods: A systematic review and a meta-analysis were conducted following the Meta-analysis of Observational Studies in Epidemiology group and Preferred Reporting Items for Systematic Reviews and Meta-analysis guidelines. Electronic databases including PubMed, Embase, and the Cochrane Library were searched for publications from inception to 15 July 2020. The inverse-variance weighted method was used in the meta-analysis to calculate the pooled odds ratios (ORs) and their 95% confidence intervals (CIs) to determine the association between snoring and MetS (and its components) through a fixed or random effect model. A restricted cubic spline regression model and the linear regression model were used in a two-stage dose–response meta-analysis to evaluate the non-linear and the linear trends between snoring frequency and MetS and its components.Results: A total of 40 studies with 966,652 participants were included in this study. The pooled ORs between snoring and MetS and its components, hypertension, hyperglycemia, low-HDL, high-triglyceride level, and abdominal obesity, were 1.61 , 1.23 , 1.05 , 1.09 , 1.08 , and 1.75 , respectively. Non-linear trends were detected in the five associations except for low-HDL. A linear trend was detected in the association of snoring with hypertension, hyperglycemia, low-HDL, or abdominal obesity, with ORs of 1.07 , 1.05 , 1.03 , and 1.17 , respectively.Conclusion: Snoring was a risk factor of MetS, and a dose–response relationship existed between the two. Timely intervention in identifying snorers can minimize as much as possible the risk of metabolic syndrome in those who snore. Metabolic syndrome (MetS), a combination of abdominal obesity and abnormal blood pressure, lipid, and/or glycometabolism , is an eIn view of such a situation, Xiong et al.  retrieveIn this study, we performed a systematic review on the relationships between snoring and MetS and each of its five components. For studies reporting snoring as a dichotomous variable, we performed a typical meta-analysis, while a dose–response meta-analysis was performed for studies reporting snoring as a ranked variable in order to give more convincing conclusions.The present study was conducted in accordance with the guidelines of the Meta-analysis of Observational Studies in Epidemiology group  and PrefIn this study, electronic databases  were searched for publications prior to 15 July 2020. The following combinations of medical subject headings and text words were used to retrieve eligible studies: ((((((((((((((MS) or Mets) or X syndrome) or syndrome X) or metabolic syndrome) or metabolic syndrome X) or insulin resistance syndrome))) OR obesity) OR (((hypertension) or high blood pressure))) OR ((((((diabetes) or diabetic) or hyperglycemia) or diabetes mellitus) or high blood sugar))) OR (((((((((triglyceride) or hypertriglyceridemia) or HDL) or cholesterol) or hypercholesterolemia) or hyperlipidemia) or dyslipidemia))) OR (((cardiovascular risk) or cardiometabolic risk))) AND (snoring or snorer). We carefully read degree theses and conference papers before excluding them from this study and contacted the corresponding authors to obtain more data for the related studies that did not provide the information that our study needed. Meanwhile, we also retrieved potential records by reviewing the references in eligible studies to cover all related literatures as completely as possible.(1) Population: adults (≥18 years) except for pregnant women; (2) interventions/exposures and controls: snoring, data for which were obtained by questionnaire in the studies. If the answer was dichotomous, i.e., “yes” or “no,” the participants who reported “yes” were defined as snorers, else as non-snorers. If the answers were ranked in different frequency levels, the participants who reported “never” were defined as non-snorers, else as snorers (more details can be seen in The inclusion criteria were as follows: (1) studies exploring the association between snoring and MetS or its components, (2) observational studies including cross-sectional, case–control, or cohort studies, and (3) ORs (95% CIs) were provided or could be calculated using relevant information.The exclusion criteria were as follows: (1) animal studies, (2) notes, case reports, comments, guidelines, reviews, letters, conference abstracts, no full text published later, and meta-analysis, (3) not in English, (4-6) studies including children, adolescents, or pregnant participants, (7) obesity defined by body mass index (BMI) rather than abdominal obesity defined by waist circumference, (8) endpoints of type I diabetes, (9) ORs (95% CIs) were reported between different levels of snoring frequency and endpoints, but no information for the number of cases at each level was provided, and (10) duplicate publications.Data were extracted by two authors (JM and HZ) independently. Where a disagreement occurred, a third person made the judgment. The extracted information included publication year, first author, study type, sample size, proportion of women, age, ORs (95% CIs), definition of snoring, definition of end-points, and adjusted confounders. For the ORs (95% CIs), if a study reported different values for various adjustments of covariates, the most fully adjusted ORs (95% CIs) were extracted in the final analysis. We also recorded the number of participants as well as the number of those with MetS at each level of snoring to ensure a successful dose–response meta-analysis.The quality assessment of the studies was performed by two authors (JM and HZ) independently. Case–control studies and cohort studies were assessed according to the Newcastle–Ottawa Scale , with scvia the inverse variance method, where OR > 1 showed that snoring was a risk factor of MetS, whereas OR <1 indicated that snoring was a protective factor. If the 95% CI of the pooled OR included the null value, that is 1, no evidence was found to support the statistical association between snoring and MetS. A heterogeneity assessment was completed using the Cochrane Q-test and I2 index. P < 0.10 or I2 > 50% implied a high heterogeneity among the included studies, and a random effect model was used. Otherwise, a fixed effect model was used. Meta-regression and subgroup analysis were performed to search for the heterogeneity sources and to compare them among groups. In the sensitivity analysis, one study at a time was deleted in order to assess the influence of the deleted study on the pooled results. Begg's rank correlation test  were used to evaluate the relationships between snoring and MetS (and its components) ion test , Egger'sion test , and a fp < 0.05 signified a statistically significant non-linear or linear dose–response relationship.In a two-stage dose–response meta-analysis, the dose in each level of snoring frequency was defined by the median of the corresponding range. For instance, if snoring 5–7 days per week was defined as habitual snoring in the original study, the dose for habitual snoring in the dose–response meta-analysis was 6 days per week. For studies that only classified snoring frequency into “never,” “occasional,” and “regular” but no specific explanation was provided, we define their doses as “0 day/week”, “1–3 days/week”, and “4–7 days/week”, respectively. A restricted cubic spline regression model with four knots at fixed percentiles  of the distribution was performed to evaluate the non-linear trend. A linear regression model was performed to evaluate the linear trend. The non-linear and the linear dose–response trends were tested by a chi-square test, and n = 8), Europe (n = 11), and Asia (n = 21). The process of study screening is shown in A total of 5,515 potential studies were initially searched, including 1,927 records from PubMed, 3,540 from Embase, and 48 from the Cochrane Library. Ultimately, 40 studies involving 966,652 participants were reserved for the analysis after screening according to the inclusion and the exclusion criteria: 10 cohort studies, two case–control studies, and 29 cross-sectional studies. In one study, both cross-sectional and cohort study designs were used. Among these studies, 15 cross-sectional studies, two case–control studies, and seven cohort studies were classified as high-quality studies. Fourteen cross-sectional studies were classified as medium-quality studies. Three cohort studies were classified as low-quality studies see –3. The sI2 = 45.30%, p = 0.08), and the pooled OR was 1.61   see .In the sensitivity analysis, when studies were deleted, one at a time, the ORs changed from 1.87  to 2.05 , as shown in p-value for Egger's test was 0.02, implying bias among included studies, the p-value for Begg's test was 0.17, implying no bias. The funnel plot can be seen in Although the Q-test result indicated the presence of heterogeneity  among the six studies. Following the use of the non-linear and the linear random effect dose–response models, the hypothesis testing for a non-linear trend was found to be statistically significant ; thus, the fixed effect model was used for further analysis. The pooled OR was 1.23  to 1.27 , as shown in p = 0.37 for Begg's test and p = 0.87 for Egger's test), and the funnel plot was symmetrical ; therefore, the random effect model was used. Both non-linear and linear dose–response models showed statistically significant results . The non-linear curve showed ORs of 1.07  for 0.5 day/week, 1.21  for 1.5 days/week, 1.26  for 2 days/week, 1.31  for 3 days/week, 1.32  for 3.5 days/week, 1.34  for 4.5 days/week, 1.36  for 5 days/week, 1.38  for 5.5 days/week, 1.44  for 6.5 days/week, and 1.47  for 7 days/week   see .The sensitivity analysis showed stable pooled results ranging from 1.05  to 1.10 , as shown in p-value for Egger's linear test was 0.01, implying the existence of bias among the studies. However, the p-value for Begg's rank was 0.56, indicating no bias. The funnel plot can be seen in The 2 = 155.56, p < 0.001); thus, a random effect model was performed. Both the non-linear and the linear dose–response models were statistically significant . The non-linear model displayed ORs of 1.04  for 0.5 day/week, 1.12  for 1.5 days/week, 1.15  for 2 days/week, 1.21  for 3.5 days/week, 1.25  for 4.5 days/week, 1.28  for 5 days/week, 1.32  for 5.5 days/week, 1.36  for 6 days/week, 1.40  for 6.5 days/week, and 1.45  for 7 days/week   see .As shown in p-values were 0.94 and 0.41 for Begg's test and Egger's tests, respectively; thus, the funnel plot was symmetrical as shown in The 2 value for heterogeneity was 106.48 (p < 0.001); thus, a random effect model was used. There were non-linear and linear dose–response trends of the association between snoring and dyslipidemia . The non-linear model showed ORs of 1.21  for 1.5 days/week, 1.26  for 2 days/week, 1.42  for 4.5 days/week, 1.46  for 5 days/week, and 1.60  for 6.5 days/week  among “studies reporting on HDL,” implying that a fixed effect model was needed. There were linear trends but no non-linear trend between snoring and HDL  among “studies reporting on triglyceride,” implying that a random effect model was needed. There were non-linear but no linear dose–response relationships between snoring frequency and triglyceride , with 5,614 participants. The pooled OR (95% CI) was 1.75 , as shown in There was heterogeneity among four studies , 11, 22 The sensitivity analysis showed that the ORs ranged from 1.47  to 1.89  by deleting the included studies one at a time .p-values for Egger's linear test and Begg's rank test were 0.296 and 0.387, respectively, which indicated that there was no publication bias and a symmetrical funnel plot was found (see The ound see .2 for heterogeneity was 5.27 (p = 0.38); hence, a fixed effect model was used. There were non-linear and linear dose–response trends in the relationship between snoring and abdominal obesity . The monotonous curvilinear association showed ORs of 1.63  for 1.5 days/week, 1.77  for 2 days/week, 1.95  for 4.5 days/week, 2.06  for 5 days/week, and 2.74  for 6.5 days/week, as shown in In three studies from two articles , 26 withThis meta-analysis of one cohort study and seven cross-sectional studies displayed a statistically significant linkage between snoring and MetS. Furthermore, the dose–response meta-analysis of six other studies revealed a non-linear trend of the role of snoring on MetS. Our previous study revealed a similar association even when the interaction effect between snoring and obesity was considered on the prevalence of MetS . The stuThis study showed that snoring was a risk factor for hypertension both in pooled results and in the dose–response meta-analysis. In the subgroup analysis of different study designs, case–control studies showed the opposite results of a non-significant OR of 1.25 . The two case–control studies were both by Marrone et al. , for botAs mentioned in the introduction, Xiong et al. performed a meta-analysis on the relationship between snoring and diabetes in 2016 , but onl3, but not in those with a BMI ≤ 25kg/m3 in a multivariable analysis. Furthermore, Huang et al. .For an association between snoring and dyslipidemia, Cho et al.  showed tg et al.  conclude2, but not in participants with a BMI <25 kg/m2. Similar studies (The present study showed a significant association between snoring and abdominal obesity in typical and dose–response meta-analysis. Previous studies have reported obesity as a major cause of snoring , whereas studies , 43 also studies . TherefoBetween-study heterogeneity is common in meta-analysis. The meta-regression results showed that the different definition criteria for MetS are associated with heterogeneity in the meta-analysis for MetS . The stuTo the best of our knowledge, this is the first comprehensive and systematic review to reveal the associations between snoring frequency and MetS and its components. Meanwhile, our analysis provides conclusion on the dose–response trend between snoring and MetS, which provided support for further research on the causal association between snoring frequency and MetS. However, several limitations should be noted. Firstly, the risk factors for MetS still remain unclear; therefore, the confounding variables in various studies differed. However, we extracted the ORs with most adjusted factors in our analysis. Secondly, most studies included in our analysis were cross-sectional or case–control studies that did not incorporate time sequence criteria for determining causality. Furthermore, only one study  excludedIn conclusion, we observed significant associations between snoring frequency and MetS and its components both in typical and dose–response meta-analyses. Our findings, therefore, have implications for metabolic dysfunction management. Early detection of snoring and early intervention in snorers may decrease the risk of MetS. Because of the intrinsic limitations of the included studies, more prospective cohort studies with participants who are simple snorers, thus excluding patients with OSA, are still needed to establish the potential causal relationship between snoring and MetS.TW was responsible for the study concept and design. JM and HZ completed the study selection and study evaluation. HW helped to perform the dose–response meta-analysis. JM performed the information extraction and wrote the manuscript. QG, HS, SH, LM, and TW reviewed and edited the manuscript. All authors read and approved the final 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.