Atrial fibrillation (AF) is associated with adverse events after cardiac surgery. Multiple studies have reported that posterior pericardiotomy (PP) may be effective for preventing AF after coronary artery bypass grafting (CABG), but some conflicting results have been reported and the quality of evidence from previous meta-analyses has been limited. The present study aimed to systematically evaluate the safety and efficacy of PP for preventing AF after CABG in adults.We conducted a quantitative meta-analysis of randomized controlled trials (RCTs) published before May 31, 2021. The primary outcome was AF after CABG under cardiopulmonary bypass. Secondary outcomes included early pericardial effusion, late pericardial effusion, pericardial tamponade, pleural effusion, length of hospital stay, length of intensive care unit (ICU) stay, pulmonary complications, intra-aortic balloon pump use, revision surgery for bleeding, and mortality.P < 0.0001). PP also effectively reduced the post-CABG occurrence of early pericardial effusion , late pericardial effusion , and pericardial tamponade  as well as the length of ICU stay , while increasing the occurrence pleural effusion . No significant differences length of hospital stay , pulmonary complications , revision surgery for bleeding , use of IABP , or death  were observed between the PP and control groups.Ten RCTs with 1829 patients (910 in the PP group and 919 in the control group) were included in the current meta-analysis. The incidence of AF was 10.3% (94/910) in the PP group and 25.7% (236/919) in the control group. A random-effects model indicated that incidence of AF after CABG significantly lower in the PP group than in the control group (risk ratio = 0.45, 95% confidence interval 0.29–0.64, PP may be a safe, effective, and economical method for preventing AF after CABG in adult patients.The online version contains supplementary material available at 10.1186/s13019-021-01611-x. Postoperative atrial fibrillation (POAF) is one of the most common complications after cardiac surgery –3. The tWe performed this meta-analysis in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analysis (PRISMA) statement . BecauseThe search strategy was based on the Cochrane System Review Manual . Using tThe following inclusion criteria were used: (1) study design of randomized controlled trial (RCT); (2) study population of adult patients (≥ 18 years old) undergoing CABG surgery under cardiopulmonary bypass (CPB); (3) intervention based on random assignment to receive PP or conventional treatment (no PP); (4) comparison of PP group to control group (no PP); and (5) outcome measurement related to occurrence of AF.The exclusion criteria were: treatment with surgery other than CABG, treatment without CPB, a non-randomized design, an animal study, and patients age < 18 years.All data were independently extracted by two researchers (M.Y.F and Z.Y.L) in duplicate. The following data were independently extracted using standardized and pilot data spreadsheets: first author, year of publication, type of surgery, study design, sample size, patient characteristics , intraoperative data, length of stay, length of ICU stay, AF, early pericardial effusion, late pericardial effusion, pleural effusion, pulmonary complications, intra-aortic ballon pump (IABP) use, pericardial tamponade, revision surgery for bleeding, death in the PP group, and number of individuals in the control group. The extracted data were checked by another investigator, and any inconsistencies were resolved through consensus and discussion. The primary outcome was the occurrence of AF.The quality of evidence provided by each included study was evaluated using the recommended Cochrane bias risk assessment tool. The tool evaluates randomization (selection bias), allocation concealment (selection bias), blinding of research participants and personnel (performance bias), blinding of outcome assessment (detection bias), incomplete outcome data (attrition bias), selective reporting (reporting bias), and blinding of non-results information (other bias). Each study was assessed as having a low, unclear, or high risk of bias.The Grades of Recommendations Assessment Development and Evaluation (GRADE) scale was used to assess the level and strength of evidence for recommendations, as follows: high quality: further research is unlikely to change our confidence in the estimation of effects; moderate quality: further research may affect our confidence in the effect estimate and may change the estimate; low quality: further research is likely to have an important impact on our confidence in the effect estimate and may change the estimate; and very low quality: we are very uncertain about this estimate.To avoid the increased risk of Type I errors in this meta-analysis due to the scarcity of data and repeated cumulative data testing, TSA was used to determine whether evidence was reliable and conclusive. When the cumulative z-curve crosses the test sequence monitoring boundary or an invalid area, the level of evidence for intervention is sufficient and no further testing is required. If the z-curve does not cross any boundaries, there is not enough evidence to draw a conclusion, indicating that further research is still needed. In the current study, we used an alpha error of 0.05, a beta error of 0.20, a 20% reduction in the expected risk ratio (RR) of POAF, and the proportion of events from the control group in our meta-analysis to calculate the information sample size required for TSA.We chose the main outcome to be the occurrence of POAF. Secondary outcomes included in-hospital or out-of-hospital mortality, pulmonary complications, early pericardial effusion, late pericardial effusion, pericardial tamponade, pleural effusion, length of stay in the hospital, length of stay in the ICU, use of IABP, and revision surgery for bleeding.2 and χ2 tests to test the heterogeneity between test results. Statistical heterogeneity was determined by evaluating the I2 value [For each basic hypothesis, the statistical significance level of the two-tailed test was 0.05. All statistical analyses were performed using Review Manager version 5.4  and STATA version 14 . The results are expressed with the Mantel–Haenszel RR and 95% confidence interval (CI)  . We applI2 value , accordiI2 value . All metI2 value , 31 wereFigure The characteristics of the 10 included RCTs and their participants are presented in Table 2 = 64%, Q-test P = 0.003, and effect size RR = 0.40 [95% CI 0.32–0.50, P < 0.05) indicated heterogeneity among the selected studies. On sensitivity analysis, further exclusion of any single study did not substantially change the overall combined RR, which ranged from 0.37 (95% CI 0.29–0.48) to 0.47 (95% CI 0.37–0.59). The random effects model combined with RR showed that PP significantly reduced the incidence of AF after CABG . As shown in Fig. A total of 1829 participants were included in the present analysis (910 in the PP group and 919 in the control group). The cumulative incidence of AF was 10.3% in the PP group and 25.7% in the control group –35. Fixe2 = 64%, P = 0.004; Additional file 2 = 64%, P = 0.004; RR = 0.45; 95% CI 0.31–0.67, P < 0.0001), which were statistically significant and provided substantial evidence of heterogeneity. In addition, this analysis may be affected by the use of β-blockers. Four studies in which patients did not take β-blockers before surgery were analyzed and no significant heterogeneity was observed , with statistical significance [2 = 40%, P = 0.14; Additional file P < 0.05) [We next tested the source of heterogeneity via a sensitivity analysis. Upon exclusion of the study with the smallest sample size , the fixificance –21. As m < 0.05) , 34.P < 0.05; Additional file P < 0.05; Additional file P < 0.05; Additional file P < 0.05; Additional file P < 0.05; Additional file P = 0.66; Additional file P = 0.97; Additional file P = 0.60; Additional file P = 1.0; Additional file P = 0.41; Additional file Compared with the control treatment, PP effectively reduced the post-CABG occurrence of early pericardial effusion , which further clarified the efficacy of PP for preventing AF after CABG. At the same time, in this study, because the included studies were all RCTs, clinical heterogeneity was the main source of statistical heterogeneity, with studies in Turkey accounting for 60% of all studies. To further exclude clinical heterogeneity, we performed a meta-analysis of the six prospective RCTs from Turkey, and the results again showed that PP reduced the incidence of AF after CABG . While these findings may indicate that PP can improve the incidence of AF after CABG, we cannot rule out the effects of potential bias in the included studies.While the previous meta-analyses reported that PP significantly reduced the incidence of AF after cardiac surgery , 25, thoP < 0.05) and the incidence of advanced pericardial effusion  were significantly lower in the PP group than in the control group, indicating that PP can significantly reduce the incidence of pericardial effusion in patients after CABG. Previous research has shown that the incidence of supraventricular arrhythmias is higher in patients with pericardial effusion than in those without pericardial effusion [Our analysis of the effects of PP on early pericardial effusion, late pericardial effusion, pleural effusion, pulmonary complications, length of hospital stay, length of ICU stay, IABP use, and mortality produced varying results. With regard to postoperative pericardial effusion, our study showed that the incidence of early pericardial effusion . In addition, the incidence of pleural effusion in the PP group was significantly higher than that in the control group , indicating that through the PP process, fluid can be discharged freely into the left thoracic cavity, thereby significantly reducing the incidence of pericardial tamponade [P = 0.97). Therefore, the present study indicates that PP provides an effective method for chest drainage to reduce the occurrence of cardiac tamponade without increasing the risk of pulmonary complications. More trials are needed to confirm this finding.The present meta-analysis showed that PP effectively reduced the incidence of postoperative cardiac tamponade compared with that in the control group . Because the duration of treatment in the ICU is related to total hospitalization expenses [P = 0.66), IABP use , need for revision surgery for bleeding , or mortality  between the PP and control groups. Although we found no correlation between PP and these outcomes, these events were very infrequent, which precluded a comprehensive safety assessment of these results. While these results may indicate that the PP procedure did not affect hospital stay, lung complications, revision surgery for bleeding, use of IABP, and mortality, this was an observational analysis that could produce misleading results. Therefore, these results should be interpreted with caution.Our analysis of the postoperative ICU stay showed that PP shortened the ICU stay compared with that in the control group  after removal of the study with the smallest sample size. Figure S2. Pooled analysis for the comparison of the risk for postoperative atrial fibrillation (POAF) without preoperative oral β-blockers. Figure S3. Pooled analysis for the comparison of the risk for postoperative atrial fibrillation (POAF) in Turkey. Figure S4. Pooled analysis for the comparison of the risk for postoperative early pericardial effusion. Figure S5. Pooled analysis for the comparison of the risk for postoperative late pericardial effusion. Figure S6. Pooled analysis for the comparison of the risk for postoperative pericardial tamponade. Figure S7. Pooled analysis for the comparison of the risk for postoperative length of stay in intensive care unit (ICU). Figure S8. Pooled analysis for the comparison of the risk for postoperative pleural effusion. Figure S9. Pooled analysis for the comparison of the risk for postoperative length of hospitalization. Figure S10. Pooled analysis for the comparison of the risk for postoperative pulmonary complications. Figure S11. Pooled analysis for the comparison of the risk for postoperative revision for bleeding. Figure S12. Pooled analysis for the comparison of the risk for postoperative intra-aortic balloon pump (IABP) usage. Figure S13. Pooled analysis for the comparison of the risk for postoperative death. Table S1. Main postoperative data from random controlled trials included in the meta-analysis. Table S2. Actual mode of PP and use of posterior pericardial drains.


Several studies published to date have shown inconclusive results in the association between serum bilirubin and risk of site-specific cancer types and overall cancer. Therefore, there is a need to further investigate this association. Data from the large Swedish Apolipoprotein Mortality Risk (AMORIS) cohort study was used. We found that overall high levels of bilirubin had no association with overall cancer risk. However, a positive association was found between melanoma and breast cancer risk. On the other hand, an inverse association was found between high levels of bilirubin and risk of gynecological and lung cancers. Further studies are required to establish if bilirubin can be used as a biomarker for risk assessment and/or as a novel therapeutic target.Bilirubin has strong antioxidant properties that have been hypothesized to be preventive against the development of cancer. Thus, we aimed to investigate the association between serum total bilirubin (STB) and risk of overall and site-specific cancers in the large Swedish Apolipoprotein Mortality Risk (AMORIS) cohort. We also performed a systematic review and meta-analysis for specific cancer types . We found no association between high levels of STB and risk of overall cancer. Regarding site-specific cancer, there was an inverse association between increased STB and lung cancer (Hazard Ratio (HR) for the 4th quartile (Q4) vs. Q1: 0.50; 95%CI: 0.44–0.59) and gynecological cancer . A positive association was found with melanoma  and breast cancer  risk. The meta-analysis showed an inverse association between high levels of STB and risk of lung cancer (Relative risk (RR): 0.69; 95%CI: 0.55–0.86). No associations were seen for colorectal and breast cancer risk. Further studies are required to establish if bilirubin can be used as a biomarker for risk assessment and/or as a novel therapeutic target. It has been proposed that endogenous bile pigments, such as bilirubin, may play a protective role against the development of a variety of chronic diseases, including cancer, due to their potent anti-inflammatory and antioxidant properties . Serum tSTB is commonly used as a biomarker of hepatobiliary and hematopoietic diseases . Of latePrevious prospective studies looking at the association between pre-diagnostic levels of STB and site-specific cancer risk have revealed mixed results ,13,14. MIn the current study, we aimed to analyze pre-diagnostic levels of STB in relation to the development of overall, as well as site-specific, cancer using data from the Swedish Apolipoprotein Mortality Risk (AMORIS) cohort. Additionally, we performed a systematic review and meta-analysis of the existing literature to further evaluate these associations.Our study population consisted of participants enrolled in the AMORIS cohort, with over 800,000 men and women, mainly from the greater Stockholm area, who underwent health examinations between 1985 and 1996 . They weEligible participants in the current study were men and women, aged 20 years or older, who had no record of prior cancers in the National Cancer Register. Subjects with no baseline measurement of serum bilirubin available were excluded from this study. Participants diagnosed with cancer within the first year of study commencement were also excluded, to account for reverse causation, resulting in a final study sample of 137,045 men and women. In addition to overall cancer, we also investigated major cancer types . These included: colorectal (CRC), lung (LC), breast (BC), gynecological (GYN), prostate, bladder, melanoma and hematological cancers.For the purpose of this study, we used the following information: age at measurement, serum total bilirubin , fasting status, body mass index (BMI), socioeconomic status (SES), education level, Charlson Comorbidity Index (CCI), cancer diagnosis, death and emigration dates. STB was measured within bile pigment using a diazo reagent to produce azobilirubin. Multivariate Cox proportional hazards models were used to calculate hazard ratios (HRs) and 95% confidence intervals (CI) for overall and site-specific cancer risk. Follow-up time was defined from date of bilirubin measurement until date of cancer diagnosis, death from any cause, emigration or end of study (31 December 2012), whichever occurred first.For the analyses, STB was examined both as a continuous variable and a categorical one (using quartiles). Guided by a directed acyclic graph (DAG) to assess relationships between potential confounders , we deveThe current study was conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) guidelines. No protocol has been registered for the study. A literature search of epidemiological studies was conducted on 15 May 2021 using the search engine PubMed, a bibliographic database including over 30 million references to journal articles, with the following search terms: (“Neoplasms”[Mesh] OR “cancer risk”[Title/Abstract]) AND (“Hyperbilirubinemia”[Mesh] OR “bilirubin”[Mesh] OR “bilirubin levels”[Title/Abstract]). The search was restricted to human studies published from January 1955 to May 2021. The inclusion criteria considered studies on adults only. No restrictions were placed on publication type. Non-English publications, duplicate studies, preprints, errata and animal studies were excluded. Moreover, only publications with full text available were included.Publications were initially screened by title and abstract, with potentially relevant studies undergoing a full-text review. Following the inclusion and exclusion criteria described above, some studies were further excluded in the full-text review. No publications on survival or therapeutics were included.For our meta-analysis, the following set of inclusion criteria was used to select the final set of studies: the publication pertained to an epidemiological study, which measured STB levels in association with any site-specific cancer risk; the type of cancer was defined as well as the measurement and values of STB; the analytical methods were all described; the study was performed on the whole population . To include studies of large enough power, only those with at least 20 cancer cases were included. Initially, in the screening phase, titles and abstracts of articles were reviewed to ascertain whether they might potentially fit the inclusion criteria. If, after assessing the abstract, there was any doubt over whether it met the relevant criteria, it was subjected to a subsequent assessment. The list of potential articles was further shortened by performing a detailed evaluation of the methods and results of each of the remaining papers. In addition, we included results from the analysis we performed in the AMORIS cohort. The following details were recorded for each study included in our meta-analysis: author, year of publication, country where the study was undertaken, study type, population, exposure variable, outcome , findings and adjustments.The effect of STB on site-specific cancer risk was evaluated by calculating the random effects summary relative risk to allow for possible heterogeneity between study results. The effect was only analyzed for CRC, breast and lung cancers as our search did not find enough studies for other tumor types that met the inclusion criteria. If a study reported both clinical cut-off and Q4, Q4 was used. Studies using STB continuous values were also included. Potential heterogeneity of the study results was assessed with a forest plot. Potential heterogeneity of the study results was also statistically evaluated using the I2 statistic. All analyses were performed using STATA, version 16 .n = 21,963, HR for Q4 vs. Q1: 0.45; 95%CI 0.24–0.86), while no significant associations were found in smokers  resulted in 45 articles. Eighteen articles were found looking at site-specific cancers . After ep = 0.000), which can also be observed in the corresponding forest plot (p = 0.000) also suggested heterogeneity  of 0.69 (95%CI: 0.0.33–1.46). The I2 statistic detected substantial heterogeneity between studies , found a positive association between bilirubin levels and risk of CRC cancer in men, and an inverse association in women. In addition, the authors performed a Mendelian randomization analysis of the main UGT1A1 single-nucleotide polymorphism (SNP) associated with Gilbert’s Syndrome. These analyses reported a positive association between high bilirubin and CRC risk in men, while there was no association in women ,12,14,30The strengths of our cohort study include the cohort size with 19 years of follow-up (on average) and reliable outcome data prospectively collected through linkages with national registries. New cases of cancer were identified using the Swedish National Cancer Register with a mandatory reporting of diagnosed cases. Furthermore, STB was measured at the same clinical laboratory for all subjects using a consistently implemented and well-documented methodology. One limitation of our cohort study was that the exposure variables were measured only once (at baseline), so potential changes over time could not be accounted for. Various confounding factors could not be included in our analysis due to the low percentage of participants with measurements . In addition to needing more data on smoking status, a more detailed account of smoking characteristics  would have been useful to perform further stratified analyses in the association between STB and lung cancer risk, as the current study only examined never vs. ever smokers (no data on smoking intensity). In addition, measures on UCB were not included in the AMORIS cohort. Finally, given the homogeneous ethnic background of our cohort population , we were limited in our exploration of other ethnic groups. For the meta-analysis, we made all possible efforts to include all relevant publications available to date through PubMed, one of the main online databases. However, one major limitation was the low number of studies fulfilling our inclusion criteria. In addition, there was high heterogeneity between studies, which may have been due to differences in study design. Future studies should further explore the association between STB and risk of cancer, in particular risk of lung cancer, taking a better look into the role smoking status plays in the association.Our findings suggest no association between STB levels and risk of overall cancer. However, a consistent inverse association was found between high levels of STB and risk of lung and gynecological cancer. In addition, a positive association was found with risk of melanoma and breast cancer. The results from our meta-analysis supported the association between STB levels and lung cancer, while no association was found with risk of CRC or breast cancer. Additional studies analyzing the association between STB and site-specific cancers are needed—including preclinical studies and those with detailed information on smoking.


This study aims to systematically review and meta-analyze the association between PNPLA3 2 index and using Comprehensive Meta-Analysis software.Following a systematic review and meta-analysis method, articles without any time limitation, were extracted from SID, MagIran, IranDoc, Scopus, Embase, Web of Science (WoS), PubMed and ScienceDirect international databases. Random effects model was used for analysis, and heterogeneity of studies was investigated considering the IThe odds ratio of CC genotype in patients with non-alcoholic fatty liver demonstrates the protective effect of CC genotype with the ratio of 0.52, whereas CG genotype presents an increasing effect of CG genotype with the ratio of 0.19, and GG genotype also showed an increasing effect of GG genotype with the ratio of 1.05. Moreover, CG + GG genotypes as a single group demostrated an odds rartio of 0.88.This meta-analysis highlights that people with CC genotype has 52% lower chance of developing non-alcoholic fatty liver disease, and those with CG genotype had 19% higher risk of developing non-alcoholic fatty liver. Those with GG genotype were 105% more likely to develop non-alcoholic fatty liver than others. Moreover, those present in a population with CG + GG genotypes were 88% more likely to have non-alcoholic fatty liver disease. Non-alcoholic Fatty Liver Disease (NAFLD) is a common disorder that is known to be the leading cause of chronic liver disease worldwide. This disorder is caused by abnormal accumulation of fat in liver tissue cells and can eventually lead to liver cirrhosis , 2. The In people with NAFLD, deaths from liver disease were reported to be 0.77 per thousand people per year, and cardiovascular deaths were 4.79 per thousand people per year . AccordiThe prevalence of NAFLD is associated with disorders such as obesity and insulin-resistant diabetes and metabolic syndrome , 8. In trs738409) in the patatin-like phospholipase domain containing protein 3 (PNPLA3). This SNP was first identified in 2008 by two independent studies on the independent genome.Different types of genes may be involved in the pathogenesis of NAFLD. Genetic factors cause NAFLD in 27 to 39% of cases. One of the most important genetic factors for NAFLD is Single Nucleotide Polymorphism (SNP) .The PNPLA3 rs738409 is susceptible to NAFLD.PNPLA3 encodes a protein known as adiponutrin (ADPN). This protein is expressed in adipocytes and hepatocytes. Moreover, this protein has lipolytic and lipogenic properties, however the exact function of adiponutrin is still unclear. PNPLA has also been reported to be highly expressed on human stellate cells. The encoded protein has retinol esterase activity and allows retinol secretion from hepatocytes while the mutation induces intracellular retention of this compound, therefore, PNPLA3 rs738409 is still unknown, however in vitro studies have shown that PNPLA3 protein has tricylglycerol (TG) hydrolase and lysophosphatidyl acyltransferase (LPAAT) and calcium independent phospholipase A2 activities. PNPLA3 also plays a critical role in homestasis of lipid metabolism. PNPLA3 eventually causes glycerolipid hydrolasis in the liver and inhibits lipid outflow into peripheral adipose tissue, thus contributing to hepatic steatosis and related disorders. NAFLD is characterised by the accumulation of lipids in hepatic steatosis.The function of PNPLA3 rs738409 of the patatin-like phospholipase domain containing gene 3 (PNPLA3) is known to be the most common and most potent gene in the development of NAFLD [The PNPLA3 gene is associated not only with liver fat content, but also with hepatic inflammation, hepatic steatohepatitis, fibrosis and cirrhosis, indicating that it plays a key role in the development of NAFLD. Inflammatory infiltration and liver damage are greater in patients carrying PNPLA3 I148M than in wild-type genotype individuals; this gene is thought to be closely linked to liver inflammation. Compared to non-carriers, homozygous carriers has 73% higher liver fat content, 3.2 times higher risk in high necroinflammatory scores and 3.2 times higher risk of developing fibrosis –19. rs73Furthermore, the association of PNPLA3 gene polymorphisms with other liver disorders such as alcoholic fatty liver (ALD) has also been observed ..rs738409 polymorphism and non-alcoholic fatty liver.Since non-alcoholic fatty liver disease is very common and can have adverse side effects, understanding the factors affecting its occurrence can play a key role in the prevention and control of this disease and the treatment of those affected. This study aims to systematically review and meta-analyse the association between PNPLA3 rs73409 C > G polymorphism using a systematic review and meta-analysis. Data were collected from Iranian and international databases of Web of Science (WoS), Embase, Scopus, PubMed, science direct, ProQuest, Google Scholar, SID, Irandoc and other international databases. International databses were searched using the keywords (PNPLA3 gene or PNPLA3 polymorphism OR patatin-like phospholipase domain-containing protein3) and  and their possible combination; Persian equivalent of keywords were used for searches within the Persian databases. The Google Scholar search engine was also used with both English and Persian keywords. In order to assess gray literature review, sites related to the subject, as well as the references within the found sources were analysed.This study was performed to determine the association between PNPLA3 Following the search process, all articles were collected in the EndNote software, and all duplicates were removed. Inclusion criteria were: 1- Case control studies, 2- Cohort, and 3- Studies examining the relationship between pnpla3 gene and non-alcoholic fatty liver disease and Exclusion criteria were: 1- Cross-sectional studies, 2- Case reports, 3- Intervention studies, 4- Letters to editor, 5- Studies where the full-text was not available, and 6- Studies in which individuals in the population under study have underlying disease.Then a list of titles and abstracts was prepared and after hiding the full text of the articles they were provided to the reviewers. Each article was independently reviewed by two reviewers, and in case of disagreement between the two reviewers, the third reviewer’s judgement was considered as the criterion for approval of articles.During the qualitative evaluation phase, the STROBE checklist was used to evaluate the studies qualitatively. This checklist consists of 22 criteria, of which 18 are used to assess all research papers, and 4 are specific to the type study. The checklist is used to evaluate the study objectives, determination of appropriateness of the sample size, type of study, sampling method, research population, data collection method(s), definition of variables and method of sampling, study data collection tools, study objectives, the statistical test used to assess the findings, and the maximum score derived from this checklist is 32. The articles with a score below 14 were excluded. Studies were then reviewed according to the PRISMA 2009 four-step process, including article identification, screening, eligibility criteria and finally meta-analysis.2 test, data were analyzed using Comprehensive Meta-analysis software , probability of publication bias results were evaluated using both funnel diagrams and Egger test; please note that the significance level was set at 0.05.In this study, heterogeneity of studies was investigated using Irs738409 polymorphism and non-alcoholic fatty liver disease through systematic review and meta-analysis. Following searching various databases, a total of 1391 articles entered the study, of which 220 articles were from EMBASE database, 47 articles from ProQuest, 109 articles from PubMed, 84 articles from ScienceDirect, 243 articles from Scopus, 447 articles from Web of Science (WoS), 1 article from SID, 57 articles from Irandoc, 145 articles from Google Scholar, and 2 articles were selected following the reviews of other articles, and were found within the references.This study investigated the association between PNPLA3 I148M Once the articles were collected, 360 duplicate articles were eliminated, and after reviewing the title and abstracts, 692 other articles were also removed and 339 articles were left subjected to secondary evaluation. After reviewing the full text of the articles in terms of thematic relevance as well as qualitative review of the articles, 308 additional articles were excluded and finally 31 articles entered the meta-analysis process .The heterogeneity of the studies was evaluated using the IThe total number of samples included in the case group and in the control group were 9973 and 13,048 respectively. The odds ratio of CC genotype in patients with non-alcoholic fatty liver was 0.48 based on meta-analysis (95% CI: 0.40–056), indicating an protective effect of CC genotype with 0.52, meaning that those with this genotype are 52% less likely to develop non-alcoholic fatty liver than others. In Fig. 2 test. Based on this test, I2 = 80.3% was obtained, which indicates high heterogeneity in the included studies. Moreover, the results of the publication bias study were compared with the Egger test .The heterogeneity of the studies was evaluated using the IThe total number of samples included in the case group and in the control group were 9973 and 13,048 respectively. The odds ratio of CG genotype in patients with non-alcoholic fatty liver was 1.19 based on meta-analysis (95% CI: 1–1.33), indicating an increasing effect of CG genotype with 0.19, meaning that those with this genotype are 19% more likely to develop non-alcoholic fatty liver than others. In Fig. 2 test. Based on this test, I2 = 86.3% was obtained, which indicates heterogeneity in the included studies. Moreover, the results of the publication bias study were compared with the Egger test .The heterogeneity of the studies was evaluated using the IThe total number of samples included in the case group and in the control group were 9973 and 13,048 respectively. The odds ratio of GG genotype in patients with non-alcoholic fatty liver was 2.05 based on meta-analysis (95% CI: 1.64–2.56), indicating an increasing effect of GG genotype with 1.05, meaning that those with this genotype are 105% more likely to develop non-alcoholic fatty liver than others. In Fig. 2 test. Based on this test, I2 = 90.7% was obtained, which indicates heterogeneity in the included studies. Moreover, the results of the publication bias study were compared with the Egger test .The heterogeneity of the studies was evaluated using the IThe total number of samples included in the case group and in the control group were 9973 and 13,048 respectively. The odds ratio of CG + GG genotype in patients with non-alcoholic fatty liver was 1.88 based on meta-analysis (95% CI: 1.5–2.3), indicating an increasing effect of CG + GG genotype with 0.88, meaning that those with this genotype are 88% more likely to develop non-alcoholic fatty liver than others. In Fig. rs738409 polymorphism and non-alcoholic fatty liver disease, we highlighted that people with CC genotype with the odds ratio of 0.48, have 52% lower risk of developing non-alcoholic fatty liver, while this ratio in CG and GG genotypes were 1.19 and 2.05 respectively, and therefore the probability of developing the disease in those with these genotypes were 19% (CG) and 105% (GG) higher. On the other hand, considering the CG + GG groups as a single population/group, and following a statistical analysis, it was concluded that the odds ratio of this group in relation to occurance of Non-alcoholic fatty liver was 1.88, meaning that this group were 88% more likely to develop the disorder than others. The effect of the G allele on non-alcoholic fatty liver disease can also be emphasized. A study in India in 2020 also found that the G allele plays a key role in the development of NAFLD [In this study, after investigating the association between different genotypes of PNPLA3 of NAFLD .PNPLA3 I148M and TM6SF2 E167K as the most likely genetic factors in the development of NAFLD [NAFLD is recognised as one of the most common liver diseases in the world with unknown etiology and pathogenesis. However, several factors including genetics, diet and inactivity, have been presented as some of the key reasons for the development of the disease. It has also been found that a good diet and regular exercise can reduce the risk of developing insulin resistance and can boost glucose homeostasis. Other SNPs such as rs2896019 and rs3810322 have also been reported to increase the risk of non-alcoholic fatty liver disease [of NAFLD .According to meta-analysis by Zhang et al. 2015) on some studies undertaken in Asian countries, when comparing people having the G allele with a population with the C allele, the probability of non-alcoholic fatty liver disease was reported to be 1.92, and therefore it was concluded that the G allele is likely to increase the development of non-alcoholic fatty liver to the liver in people with G allele by 92%; Moreover, it can increase the risk of renal fibrosis and ALT serum levels. Development of NAFLD in the dominant phenotype (CG + GG) was 110% higher than the recessive phenotype. On the other hand, comparing the CG + GG populations with the CG genotype, it was concluded that the risk of NAFLD was higher in the homozygous GG population than in other populations [ on some Another meta-analysis conducted in 2019 stated that this polymorphism had a major impact on the development of tissue damage in liver and that the G allele was considered as a risk factor for NAFLD in such a way that the ratio of development of the disease in those with one G allele to those without it was 1.88, and 4.01 in those where both alleles were G. It has also been suggested that this gene increases alanine aminotransferase levels in serum .According to a meta-analysis by Jiaying et al. (2020), this gene is involved in the development of non-alcoholic osteopathy (NASH) in children and adolescents; it is also accosiated with factors such as serum alanine transaminase, aspartate transaminase, gamma glutamyl transferase, that are indicators of liver damage .rs738409 polymorphism in the pnpla3 gene was strongly associated with the incidence of NAFLD and NASH, especially in Asian and Spanish populations. In this study, however, no association was found between rs738409 polymorphism and hepatic steatosis. It was also reported that the GG genotype had a high impact on the development of NAFLD as well as renal fibrosis. The ratio of this genotype over inflammation occurance was reported as 3.13 [Another meta-analysis in 2015, it was reported that all genetic variations in the  as 3.13 .According to a study by Chobin et al., the CG genotype was identified as a predisposing genotype to a 2.63-fold increase in the likelihood of developing the disease. Moreover, it was reported that the GG genotypes possess a protective effect, meaning that existence of such gentype results in a 59% decrease in developing NAFLD. Furtheremore, the odds of developing the disease in the CC genotype was 1.78 .According to another study by Sood et al. (2016) in Japan, the odds ratio of the GG genotype was 36.5% in obese people and 47.8% in the non-obese population who had a fatty liver. Moreover, the modified odds ratio of non-alcoholic fatty liver disease in GG genotype was reported to be 4.15 in non-obese individuals, and 2.76 in obese pupulation. This genotype also increases the chances of developing steatosis and liver fibrosis . A familThe limitation of this study was the lack of access to the full-text of some of the sources.This meta-analysis study demonstrated that people with the CC genotype were 52% less likely to develop non-alcoholic fatty liver disease, and people with CG genotype were 19% more likely to develop non-alcoholic fatty liver. Moreover, population with the GG genotype, had 105% more chance of developing a non-alcoholic fatty liver. Moreover, population with CG + GG genetypes demonstrate 88% more chance of developing the disease, and this is suggesting the effect of G allele on non-alcoholic fatty liver disease. In future, the effects of genetic and environmental factors on the level fo tissue damage, and also the effect of this gene on fibrosis and liver cirrhosis can be studied.


Periodontal disease is reportedly associated with the risk of various systemic diseases, including pancreatic and lung cancers. However, its association with prostate cancer remains inconclusive. Herein, we explored the association of periodontal disease with the risk of prostate cancer through a meta-analysis.MEDLINE, Embase, Web of Sciences and Cochrane Library databases were searched for eligible publications up to April 2020. Multivariate adjusted risk estimates with corresponding 95% confidence intervals (CIs) were extracted and calculated using random- or fixed-effect models.Nine cohort studies involving 3.353 prostate cancer cases with 440.911 participants were identified and included in the meta-analysis. We found that periodontal disease significantly increased the risk of prostate cancer by 1.40-fold  compared with normal condition. Interestingly, the risk of developing prostate cancer was not significant in patients treated with periodontal therapy . The results of subgroup analyses were also consistent and significant when stratified by study design and follow-up period, whereas conflicting results were observed in periodontal disease ascertainment stratification. These findings were robust as indicated by sensitivity analyses.Periodontal disease was associated with the increased risk of prostate cancer, whereas no significant association was observed in patients treated with periodontal therapy. Hence, the awareness and importance for maintaining oral health should be improved, and the underlying mechanisms linking periodontal disease and prostate cancer should be fully explored in future research. Prostate cancer is the most common cancer in men and the leading cause of cancer-related deaths worldwide , 2. ApprResults about the association between periodontal disease and prostate cancer are conflicting –13. For This systematic review and meta-analysis were conducted in line with the Cochrane Collaboration criterion  and repoMEDLINE (via PubMed), Embase, Web of Sciences and Cochrane Library databases were searched for eligible studies that investigated the association between periodontal disease and the risk of prostate cancer up to April 2020. The combination of Medical Subject Headings (MeSH) and non-MeSH terms was used in each database, including ‘periodontal disease’, ‘periodontitis’ or ‘dental health’ and ‘prostate carcinoma’, ‘prostate cancer’, ‘prostate neoplasms’ or ‘prostate tumor’, without restriction for language, region or publication status. The reference lists from previous reviews and other relevant articles were manually searched to identify additional studies. The main search was completed by the senior author (ZL Guo), and any discrepancy was resolved by consensus or consultation with another investigator (SS Wang).The inclusion criteria for original studies that investigated the association between periodontal disease and prostate cancer risk were as follows:  studies The title and abstract of all articles retrieved from the initial search were screened to ascertain their relevance. All potentially relevant full-text articles were further considered and assessed to determine their inclusion eligibility in the meta-analysis. Hereafter, two investigators (CM Gu and SY Li) independently extracted and crosschecked the following data of all the included studies through a predesigned evidence table: first author, study population, study design, country, participant characteristics (i.e. sample size and age), follow-up duration, periodontal disease ascertainment, therapy of periodontal disease, dental and smoking status, risk estimates with associated 95% CIs or sufficient raw data and corresponding adjusting confounding factors. If the information in the included studies was insufficient, the primary author was contacted to obtain and verify the data.The methodological quality of the included studies was assessed by two investigators (S Gan and Y Li) on the basis of the Newcastle-Ottawa scale (NOS) , which iThe association between periodontal disease and prostate cancer was evaluated using risk estimates, and their corresponding 95% CIs were extracted from the included studies via Stata version 15.0 . For consistent definitions, the differences amongst the various measures of risk estimates could be ignored because periodontal disease-related prostate cancer is a rare event; therefore, the ORs and SIRs were directly considered as HRs in the meta-analysis . I2 was The main characteristics of the included studies are summarized in In general, the quality of the included studies , 20–25 wThe results of meta-analysis revealed that periodontal disease significantly increased the risk of developing prostate cancer by 1.40 times , and the risk was greater than those without periodontal disease. Significant statistical heterogeneity was observed. Thus, we used a random-effect model . InteresSensitivity analysis validated the stability of our results by omitting every study. Notably, the results of meta-regression analyses indicated that the variables  could not result in heterogeneity amongst the included studies. Moreover, the adjusted R2 values from −12.07% to −4.10% revealed that these regressors slightly contributed to the explanation of the response variables.The association between periodontal disease and prostate cancer was assessed using nine pooled studies, which involved 3.353 prostate cancer cases amongst 440.911 participants. The results suggest that periodontal disease is associated with increased risk of prostate cancer. Interestingly, the risk of developing prostate cancer was not significant in patients treated with periodontal therapy. Moreover, the results of subgroup analyses were consistent and significant when stratified by study design and follow-up period, whereas conflicting results were observed in periodontal disease ascertainment stratification. The results were robust as indicated by the sensitivity analysis. However, the meta-regression failed to identify the potential confounding factors that might affect the level of heterogeneity amongst the included studies.Several studies shared conflicting results of the association between periodontal disease and prostate cancer –22. In aNotably, a prospective cohort regarding the topic was excluded from the meta-analysis because it did not meet the inclusion criteria. We found that Michaud et al. performed two prospective cohorts in 2016  and 2008A meta-analysis regarding the association between periodontitis and cancer risk was published by Corbella et al. . AlthougThe incidence of prostate cancer is increasing year by year with the improvement of prostate biopsy technology , 29. HowOverall, our study exhibited several crucial strengths. Firstly, it was the latest and most comprehensive meta-analysis regarding the association between periodontal disease and the risk of prostate cancer. Moreover, subgroup analyses stratified by study design, country, follow-up period, periodontal disease ascertainment and periodontal disease therapy were performed to determine whether these variables affected the level of heterogeneity amongst the included studies. Secondly, all the risk estimates extracted from the included studies were adjusted for confounding factors to minimize their effect on the overall results. Finally, sensitivity analysis and meta-regression further validated and reinforced the rationality and reliability of our findings.However, the meta-analysis was restricted by several limitations. Firstly, five studies used retrospective cohort design, which might miss data and result in a risk of bias. Secondly, significant heterogeneity was observed. Moreover, none of the variables that might affect the level of heterogeneity were identified although sensitivity analysis revealed the robustness of the overall results. Finally, we were unable to further investigate the association between smoking status and the degree of severity of periodontal disease because of the limited data. Therefore, future high-quality research should comprehensively address these issues.Existing evidence suggests that periodontal disease appears to be associated with an increased risk of developing prostate cancer. Interestingly, no significant association was observed in patients who underwent periodontal therapy. Hence, patients with periodontal disease, a high-risk group for developing prostate cancer, should be treated with periodontal therapy immediately. Furthermore, the awareness and importance of maintaining oral health should be improved based on the main findings. Notably, further research should fully explore the underlying mechanisms linking periodontal disease and prostate cancer.


To investigate the effect of antimalarials on cancer risk in patients with systemic lupus erythematosus (SLE).PubMed, EMBASE, Web of Science, and the Cochrane Library were searched from their inception to October 3, 2020. Relative risk (RR) with 95% confidence intervals (CI) was used to evaluate the results. Subgroup analyses were used to assess heterogeneity. A funnel plot was used to explore publication bias. STATA was applied for all analyses.I2 = 43.1%, p = .173). And the consistent result was also found in SLE from multiple centres  . On disease course- and comorbidities-matched studies, the pooled RRs were 0.69 (95% CI: 0.52–0.93) and 0.59 (95% CI: 0.46–0.75), respectively.A total of nine studies consisted of four nested case–control, two case–cohort and three cohort studies were included. The results showed that antimalarials might reduce the risk of cancer in SLE . In the subgroup analysis of four nested case–control and two case–cohort studies, the pooled RR was estimated as 0.69 (95% CI: 0.60–0.80). In four studies about hydroxychloroquine, the pooled RR was estimated as 0.70 (95% CI: 0.53–0.93). Antimalarials might reduce the risk of cancer in SLE among the Asian population  (Antimalarials might be protective factors for cancer in SLE.Hydroxychloroquine might be a protective factor for cancer in SLE patients.The first article to perform the meta-analysis of antimalarial drugs on the risk of cancer in SLE patients.Results of this meta-analysis showed that antimalarial drugs might be protective factors for cancer in SLE. Hydroxychloroquine might be a protective factor for cancer in SLE patients. Systemic lupus erythematosus (SLE) is an autoimmune disease that often occurs in women and affects multiple organs ,2. PatieAt present, corticosteroids, hydroxychloroquine (HCQ), and immunosuppressants are most commonly used in the clinical treatment of SLE. Immunoregulatory and anti-inflammatory effects of antimalarials have beneficial effects on the outcomes of SLE patients. It might improve the survival and remission rates , ameliorSome studies have reported relationships between antimalarials and cancer in SLE patients, but the results were controversial. Ruiz-Irastorza G put forward a hypothesis that antimalarial drugs might have a protective effect on cancer in SLE patients . But DeySupplementary Material 1). We systematically searched EMBASE, Web of Science, PubMed, and the Cochrane Library from their inception to the end of October 3, 2020. The search strategy included the MeSH and text words as  AND ) AND . In addition, relevant articles outside the search list were manually searched.This review was conducted by PRISMA guidelines  (SupplemStudies were eligible if they met the following criteria: (1) All included patients met the American College of Rheumatology criteria or international classification of disease criteria for the diagnosis of SLE. (2) The study provided the cancer outcomes in SLE patients who took antimalarials  and controls who did not take antimalarials . (3) The study was designed by cohort, case–cohort, or nested case–control. The study would be excluded if met following contents: (1) The outcome was cancer with precancerous lesions. (2) The included patients had cutaneous lupus erythematosus.The relevant information, including first author, research period, year of publication, type of study, study population, disease course, SLE and cancer diagnostic criteria, the number of SLE patients taking antimalarials, and the number of people who developed cancer in each group, were extracted independently by two reviewers (XBL and NWC). In these studies, disease duration was defined as the duration from the date of SLE diagnosis to that of their cancer diagnosis . When thI2) was calculated to determine heterogeneity. I2 < 50% indicated that the heterogeneity of included studies was acceptable. Factors, such as research type, region, types of antimalarials, gender, age, disease course, and comorbidities may induce heterogeneity. If included studies reported information about these factors, subgroup analyses were conducted to explore sources of heterogeneity. Sensitivity analysis was used to assess the robustness of estimates. Funnel plots, Egger’s test, and Begg’s test were applied to evaluate publication bias. p < .05 was defined as statistically significant.Excel 2019 and STATA 11.0  were used for data extraction and meta-analysis. Data was expressed by relative risk (RR) with its associated 95% confidence intervals (CI). Inverse variance with random effects models was used for data analysis. Inconsistency , and eight studies were finally included [A total of 2737 articles were searched at first from four databases. Two thousand six hundred and eighty-nine articles were excluded after deleting duplication and screening titles and abstracts. After reviewing the full text of the remaining, 40 studies were excluded     ( = .067) . In four = .365) . However = .009) .I2 = 43.1%, p = .173). And a consistent result was also found in SLE from multiple centres  . Whereas no significant difference was found among the European population    ( = .005) .I2 = 37.1%; p = .204)    ( = .204) . On stud = .490) . On stud = .153) .I2 = 43.1%, p = .153)   ( = .153) . In five = .053) .t = −0.88) and Begg’s test (Z = 0.52) indicated no significant publication bias (Results of Egger’s test (ion bias . Removinion bias . It indiTo the best of our knowledge, this is the first meta-analysis examining the risk of cancer in SLE patients taking antimalarials. Antimalarials might reduce cancer risk in SLE patients. In particular, HCQ might reduce the risk of cancer in SLE patients. In four nested case–control and two case–cohort, SLE patients in the antimalarials+ group had a lower risk of cancer than those in antimalarials− group. But the difference was not found in subgroup analysis in cohort studies.Three phase I trials of HCQ combined with vorinostat, bortezomib and temsirolimus in the treatment of cancer showed that the combination therapy had anti-tumour activity and had potential effects on the treatment of tumours . In a meThe antitumor properties of antimalarials may be related to their promotion of macrophage transformation, inhibition of autophagy, and promotion of apoptosis . HCQ andSeveral advantages were in this study. First, the data of our study was retrieved from prospective studies, including cohort studies, nested case–control studies, and case–cohort studies. The collection of exposure data in prospective studies is obtained by the investigator personally, and there is generally no recall bias, so the data is reliable. In prospective studies, because the exposure occurs before the occurrence of the disease and the causal time sequence is clear, the ability to test the aetiological hypothesis is stronger and the results are more reliable. Second, a large sample size of 14810 was included in this meta-analysis, indicating the high credibility of the results. Third, with all NOS scores ≥7, the qualities of studies included were relatively high. It demonstrated the high reliability and validity of the results.However, our study has several limitations related to the quality of data in original sources. First, we were unable to fully assess the risk of bias in certain situations because the published studies did not provide enough details. Due to the lack of necessary data on malignant tumour types and drug dosages, our results may be underestimated. Therefore, the drug dosage and the drug situation for each cancer should be reported in further studies. Second, among different research types, the results of subgroups were inconsistent. Therefore, the results should be further verified in the future. Third, our data in this study came from observational studies, and we need to be cautious when interpreting the evidence of observational studies, because the results of observational studies are more biased than the results of randomized controlled trials. Therefore, more experiments are needed to prove this conclusion in the future.The result of this meta-analysis showed that antimalarials might be protective factors for cancer in SLE. HCQ might be a protective factor for cancer in SLE patients.Click here for additional data file.


Metabolic syndrome is a cluster of cardio-metabolic risk factors and comorbidities, including central obesity, hypertension, hyperglycemia, and dyslipidemia. In addition, different studies have shown that the disturbances in circadian rhythm are connected with components of metabolic syndrome. Circadian rhythm is the central regulator of every aspect of human health and metabolism, and metabolic homeostasis is essential in regulating energy metabolism, especially in adipose tissue. Therefore, we aimed to evaluate the association of genetic variations of circadian rhythm genes with metabolic syndrome and its components by a systematic review and meta-analysis. Our findings suggest that some variants of the circadian rhythm gene might be genetic biomarkers applied to predict metabolic syndrome susceptibility.Metabolic syndrome (MetS) is a combination of cardiovascular risk factors associated with type 2 diabetes, obesity, and cardiovascular diseases. The circadian clock gene polymorphisms are very likely to participate in metabolic syndrome genesis and development. However, research findings of the association between circadian rhythm gene polymorphisms and MetS and its comorbidities are not consistent. In this study, a review of the association of circadian clock gene polymorphisms with overall MetS risk was performed. In addition, a meta-analysis was performed to clarify the association between circadian clock gene polymorphisms and MetS susceptibility based on available data. The PubMed and Scopus databases were searched for studies reporting the association between circadian rhythm gene polymorphisms  and MetS, and its comorbidities diabetes, obesity, and hypertension. Thirteen independent studies were analyzed with 17,381 subjects in total. The results revealed that the BMAL1 rs7950226 polymorphism was associated with an increased risk of MetS in the overall population. In contrast, the CLOCK rs1801260 and rs6850524 polymorphisms were not associated with MetS. This study suggests that some circadian rhythm gene polymorphisms might be associated with MetS in different populations and potentially used as predictive biomarkers for MetS. Metabolic syndrome (MetS) is a set of cardiometabolic disorders associated with cardiovascular risk factors ,2. MetabPER) and cryptochrome (CRY) genes make up the negative feedback loop. Additionally, BMAL1 stimulates the expression of RORα and γ genes (RAR-orphan receptor). Casein kinases I δ and ε (CKI δ and ε), which degrade CRY and PER proteins, also contribute to the circadian activation of the clock gene. In addition to CRY and PER proteins, REV-ERB proteins (reverse erythroblastosis virus) prevent BMAL1 expression. Molecular clocks exist in all body cells and show 24 h periodicity [Circadian rhythm is a significant regulator of every aspect of human health and metabolism, and metabolic homeostasis is vital in regulating energy metabolism, especially in adipose tissue ,7. The ciodicity ,10,11. A circadian metabolic disorder is a new risk factor for MetS. Many animal studies have investigated the relationship between circadian rhythm and metabolism. Circadian rhythm disorder has been associated with obesity, diabetes, hypertension, cardiovascular disease, and all components of MetS ,12,13. ILipids and glucose have an essential role in developing MetS because they are an integral part of metabolic pathways. Circadian rhythm has a vital function in the homeostasis of lipid and glucose linked to obesity. Thus, it has a role in the risk of developing insulin sensitivity, diabetes, and cardiovascular diseases ,16. In aMany studies that have shown an association between circadian rhythm genes and metabolic syndrome have been conducted in animal models, most commonly in mice and rats that are nocturnal animals. The results of such research are applied to diurnal humans. However, there have been some inconsistencies among studies performed to date. Thus, Cho et al. showed that REV-ERBα alpha was associated with hyperglycemia and high triglyceride levels . In contFurthermore, the modern lifestyle contributes to reduced light exposure during the day (low indoor lighting levels) and increased light exposure at night, directly affecting our circadian rhythm and health, including the onset of metabolic syndrome. Therefore, the present meta-analysis was conducted to elucidate the association between circadian clock gene polymorphisms and MetS susceptibility. Furthermore, the study might provide novel potential biomarkers for the prognosis of MetS risk. The databases (Scopus and PubMed) were searched for studies investigating the association between circadian clock gene polymorphisms and metabolic syndrome and its risk factors, such as diabetes, hypertension, and obesity until 6 July 2021. The searching strategy is presented in All retrieved articles were primarily selected for acceptable studies regarding specific criteria: cross-sectional or case-control studies and focused on the association of circadian clock gene polymorphisms with MetS risk and its comorbidities. The exclusion criteria included duplicate records, no control group, and departure from Hardy–Weinberg equilibrium in the control group. In addition, only articles written in English were considered. Two investigators  independently extracted data. Data on the author, publication year, study design, demographic characteristics, risk factors related to MetS, genotyping method, allele frequencies, distribution of genotypes, diagnostic criteria, and clinical characteristics were collected. Discrepancies were resolved by discussion. The quality of the studies evaluated by the Newcastle-Ottawa Scale (NOS) is based on the range from one to nine stars and three components: selection, comparability, and outcome ascertainment . Studies2 higher than 50% was considered significant heterogeneity. The sensitivity analysis was conducted to evaluate studies with high statistical heterogeneity by sequentially removing each study to detect pooled results’ stability and heterogeneity source. A meta-analysis was conducted if at least two independent studies for the same polymorphism were available. In addition, publication bias was assessed by Begg’s and Egger’s tests and visual assessment of the funnel plot. A two-sided p-value < 0.05 was deemed significant. The Comprehensive Meta-analysis software version 3.3.070  was used for statistical data processing. The principal outcome was the odds ratio (OR) and 95% confidence interval (95% CI) of specific polymorphism in patients with risk factors related to MetS compared to the controls under allelic, dominant, and recessive genetic models. The selection bias was estimated by computing the Hardy–Weinberg equilibrium (HWE) and frequencies of the genotypes. A random-effect model was used because the studies were performed in a wide range of settings in different populations. The considerable between-study heterogeneity is anticipated in genetic association studies. The heterogeneity was estimated using the Cochran Q-test and the I-squared statistic, and IA sum of 1828 articles was recovered through databases search after removing duplicate articles. Screening titles and abstracts excluded the 1791 articles. Furthermore, 24 articles were excluded based on the inclusion/exclusion criteria. Finally, 13 studies ,31,32,33CLOCK gene polymorphisms in 3644 participants with MetS risk factors and 3884 control participants [BMAL1 polymorphisms in 3644 participants with MetS risk factors and 3884 control participants [PER3 gene polymorphisms in 2156 participants with MetS risk factors and 2247 control participants [CRY1 [CRY2 [REV-ERBα [BMAL1 rs7950226, CLOCK rs1801260, and rs6850524, were incorporated in the final meta-analysis.In 13 studies included, 11 circadian rhythm genes were analyzed with 46 different polymorphisms. Six studies evaluated the icipants ,25,27,29icipants . Three sicipants ,33. Two Y1 [CRY2 ,25, and REV-ERBα ,32 gene REV-ERBα ,32. ManyREV-ERBα ,30,32,33BMAL1 and CLOCK polymorphisms with overall MetS was estimated by computing pooled ORs. CLOCK polymorphisms, rs1801260, and rs6850524, were not associated with MetS risk . In contrast, BMAL1 rs7950226 was linked with overall MetS risk  (The association of 07–1.54) .BMAL1 rs7950226 polymorphism could lower the risk for MetS comorbidities . In contrast, CLOCK rs6850524 polymorphism did not influence the risk for MetS  and BMAL1 gene (rs7950226 G > A). The results revealed that BMAL1 rs7950226 polymorphism was linked with MetS risk in the overall population, while CLOCK rs1801260 polymorphism was associated with some specific subgroups. Thus, it is the first time comprehensively assessing the study progress in this domain and the first meta-analysis of MetS-related circadian clock polymorphisms.In the present study, a systematic review was conducted to associate circadian clock gene polymorphisms with the overall risk of metabolic syndrome. In addition, a meta-analysis was conducted for three frequently investigated polymorphisms in CLOCK protein is included in the transcriptional control of circadian output genes and the core circadian clock. Hence, up to 10% of the human transcriptome may be under circadian regulation, and disorder in the CLOCK gene substantially influences transcription control.CLOCK rs1801260 (T3111C) polymorphism, placed in the 3′-UTR of the gene, has been extensively studied for its function in different MetS risk factors and in MetS patients on various diets. It is the first polymorphism recognized in the CLOCK gene to be linked with human MetS phenotypes [CLOCK rs1801260 polymorphism might influence its transcription by modifying mRNA durability and then participating in MetS development. Ozburn et al. showed that rs1801260 polymorphism alters the expression, function, and stability of CLOCK mRNA and consequently affects the PER2 expression [BMAL1 rs7950226 polymorphism might be a predictive biomarker for MetS risk in the overall population. Nevertheless, additional molecular analyses require elucidation of all the assumptions regarding relevant mechanisms.The enotypes . Howeverenotypes . However1801260 T11C polymenotypes , while ienotypes ,37,38. Menotypes  observedenotypes . The CLOpression . In concCLOCK rs1801260, rs6850524 polymorphism is an intron variant with three possible variations: C > G, C > A, and C > T, with C > G being the most common variant. Although studies revealed a discrepancy in the association of CLOCK rs6850524 polymorphism and MetS risk factors, this meta-analysis observed no association. However, some studies found an increased risk for hypertension [Unlike rtension  and obesrtension ,29 assocCLOCK rs6850524 polymorphism, BMAL1 rs7950226 polymorphism is an intron variant G > A and has been extensively studied for its function in different MetS risk factors. In the present study, a statistically significant association of BMAL1 rs7950226 with MetS risk was observed in the overall population  might affect melatonin function and influence susceptibility to insulin resistance and diet-dependent weight loss [As observed in this study,  obesity ,25,29,43mponents ,33. PER2diabetes ,30,32. Rpulation . Howeverpulation ,45. Moreght loss ,47. Someght loss ,49. HoweThe present study has some shortcomings. First, the database search was limited to English articles, so that some studies could have been overlooked. In addition, although the meta-analysis included a moderate number of participants, the critical studies are somewhat limited. Thus, this area requires further research at the molecular level and updated meta-analysis.CLOCK rs1801260 and rs6850524, and BMAL1 rs7950226). The results revealed that the BMAL1 rs7950226 polymorphism was linked with the risk of MetS in the overall population. In contrast, the CLOCK rs1801260 polymorphism was associated with particular subgroups, implying it might be a possible prognostic biomarker for MetS risk. Thus, the study might present new hints to identify genetic biomarkers related to miRNAs that predict MetS susceptibility.In conclusion, the association of circadian clock gene polymorphisms with the overall risk of MetS was reviewed. Moreover, a meta-analysis was conducted employing all possible data for three often studied polymorphisms among them (


Background: Aberrant homocysteine level is associated with metabolic disorders and DNA damage, which may be involved in the carcinogenesis of hormone-related cancers, but clinical results of observational studies are controversial. In this study, we investigated the causal relationships between plasma homocysteine and breast cancer (BRCA), prostate cancer (PrCa), and renal cell carcinoma (RCC) using Mendelian randomization (MR) analyses.Design and Methods: To investigate the putative causal associations between homocysteine and the aforementioned three types of cancers, a two-sample MR study was employed for the study. The primary strategy for summary data analyses was the inverse-variance-weighted (IVW) approach. In our study, the single-nucleotide polymorphisms (SNPs) excluded confounding factors through Linkage Disequilibrium (LD). Phenoscanner tests were the instrumental variants (IVs), homocysteine was the exposure, and BRCA, PrCa, and RCC were the outcomes. Single-nucleotide polymorphisms associated with homocysteine were extracted from a large genome-wide association study (GWAS) meta-analysis of European participants . Summary Statistics of BRCA were obtained from the latest and largest GWAS meta-analysis comprising of 82 studies from Breast Cancer Association Consortium (BCAC) studies, including women of European ancestry ; we obtained summary-level data from the GWAS meta-analysis of PrCa comprising 79,148 cases and 61,106 controls of European ancestry, and the dataset of RCC was a sex-specific GWAS meta-analysis comprising of two kidney cancer genome-wide scans for men  and women  of European ancestry. The MR-Egger and weight median analyses were applied for the pleiotropy test.Results: The results showed null associations between plasma homocysteine levels and overall BRCA , overall PrCa , RCC in men , and RCC in women .Conclusions: We found no putative causal associations between homocysteine and risk of BRCA, PrCa, and RCC. Homocysteine is derived from the essential amino acid methionine  and funcin vivo and induce the carcinogenesis of BRCA  biased results, limited cases, confounding factors, and reverse causation. It is interesting to note that published observational studies on the association of homocysteine, as a potential cancer biomarker, with risks of BRCA and PrCa have drawn inconsistent conclusions –24. One The flow chart of the MR study design is shown in P-value for the Hardy–Weinberg equilibrium (HWE) <1.0 × 10−6, call rate <95%] were used. General characteristics of the contributing cohorts and study population are described elsewhere  was set as the standard to ensure a close relationship between SNPs and plasma homocysteine levels. A total of 18 SNPs were obtained from a study of total homocysteine level datasets. Among them, three SNPs  were weeded out because of the linkage disequilibrium (LD) to conform the independence of selected IVs (r2 < 0.05)  . Given tBreast cancer, PrCa, and RCC were the outcomes. We selected the latest published GWAS meta-analysis that contained the most complete clinical data available, on three types of cancers, respectively. The sources of datasets are shown in Summary statistics of overall BRCA were obtained from the latest and largest GWAS meta-analysis comprising of 82 studies from the Breast Cancer Association Consortium (BCAC), including women of European ancestry  [BRCA–GWAS] . Two IllP-value for HWE < 1.0 × 10−12 in cases or P < 10−7 in controls, the call rate <95%, or with concordance <98%.For PrCa, we extracted summary-level data from the GWAS meta-analysis of Prostate Cancer Association Group to Investigate Cancer-Associated Alterations in the Genome (PRATICAL) consortium comprising of 79,148 cases and 61,106 disease-free controls of European ancestry [PrCa–GWAS] . This GWP-value for HWE <1.0 ×10–7, call rate <90%, and MAF < 0.05) were applied.The dataset of RCC was a sex-specific GWAS meta-analysis comprising of two kidney cancer genome-wide scans for men  and women  of European ancestry [RCC–GWAS] . HumanHap > 0.05) and indicates no horizontal pleiotropy. In addition, a weighted median method (WM), an effective statistical tool that only needs half of the effective SNPs, was used as proof for the IVW approach by calculating the median value of selected estimates of IVs  approach  was appls of IVs .All analyses were conducted by “Mendelian Randomization” package in R version 3.6.1.To establish putative causal associations between homocysteine concentrations and BRCA, PrCa, and RCC, we performed IVW analyses to estimate causal effects. The associations between genetic variants for plasma concentrations of homocysteine and all the outcomes are listed in P = 0.543; The outcomes suggested no evidence that homocysteine levels had causal effects on overall BRCA , and RCC in women   .p > 0.05), which indicated little horizontal pleiotropy (The MR-Egger and the weight medium methods were used to further analyze the influence of pleiotropic effects on the final outcome. A similar conclusion was drawn according to the sensitivity analyses. The MR-Egger test showed non-zero intercept values (eiotropy . Similareiotropy . All theThis two-sample MR study suggested no putative relationships between plasma homocysteine levels and the risk of BRCA, PrCa, and RCC. In addition, the levels of homocysteine were not significantly related to the risk of BRCAs with different ER expression status and RCC of different sexes.As a plasma component involved in one-carbon metabolism, homocysteine has attracted attention in diverse clinical studies. Higher plasma homocysteine levels tend to be associated with metabolic disorders, even cancer; but the conclusions drawn from previous observational studies on associations between levels of plasma homocysteine with cancer risk are controversial given that other studies found no evidence for casual associations between genetically-determined plasma homocysteine levels with risk of hormone-related cancers. In an EPIC–Varese cohort study , researcThe plasma homocysteine levels were not found to be significantly elevated in early stage cancers, as opposed to a sharp increase in their advanced stages . In addiMendelian randomization can utilize genetic instruments to analyze a large number of patient samples because of increasingly/readily more available published data from diverse GWAS datasets, which can significantly increase the accuracy thus the credibility of analyses. To the best of our knowledge, this is the first two-sample MR study that has accurately evaluated potential associations between plasma homocysteine level and cancer risk. The advantage of this MR approach is that genetic variation is a long-term and stable cause of exposure. Compared with randomized controlled trials, it is elaborated directly and is not subject to interferences by confound factors like social environment and lifestyle. In this particular study, the LD analyses were used to rule out three SNPs, and the MR-Egger intercept test and sensitivity analyses showed no pleiotropic effect of the 15 selected SNPs, which additionally increases the credibility of our study.de novo nucleotide biosynthetic pathways, are associated with risks of cancer development.However, our study is not without limitations. The data that the MR analyses utilized have been collected from earlier published available GWAS summary data that have been adjusted. Further, the grouped data did not allow us to effectively explore the associations between homocysteine levels with different stages of cancer and different hormone status of different cancers. In a broader sense, the methionine cycle is but one of the pathways of one-carbon metabolism; further studies are needed to explore whether other biochemical components involved in one-carbon-fixation pathways, such as those engaged in In conclusion, our study demonstrated no evidence of causal associations between plasma homocysteine and cancers of the BRCA, PrCa, and RCC via a two-sample MR approach. Herein, our MR study suggested that homocysteine alone might not be useful as a dynamic biomarker alone for the risks of the aforementioned three cancers.The original contributions presented in the study are included in the article/QH, YS, and YaL conceived the study design and drafted the manuscript. ZY, ZQ and JL participated in data extraction and data analysis. XJ and YiL did the data checking and analysis. All authors critically reviewed the manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Our study aimed to assess the association between serum cystatin C levels and prognosis in acute myocardial infarction (AMI) patients after coronary reconstructive surgery.We searched PubMed, Embase, and Cochrane Library up to January 21, 2022 without language restriction. Outcomes were major cardiovascular events (MACEs) and mortality. The risk ratio (RR) and 95% confidence interval (CI) were merged by random-effect models.P < 0.001) and mortality  in AMI patients after coronary revascularization. Subgroup analysis showed that the serum cystatin C levels were associated with significantly higher risk of MACEs  and mortality  in AMI patients after percutaneous coronary intervention (PCI). However, in AMI patients after coronary artery bypass surgery, there were no significantly higher risk of MACEs  and mortality . Further subgroup analysis showed that this significantly higher risk of MACEs and mortality did not change with the study sample size, study population area or study follow-up time.We included 8 studies with a total of 7,394 subjects in our meta-analysis. Our meta-analysis showed that higher-level of serum cystatin C levels were associated with higher risk of MACEs (RR = 2.52, 95% CI 1.63–3.89, The meta-analysis demonstrated that higher serum cystatin C levels were associated with significantly higher risk of MACEs and mortality in AMI patients after PCI. It is a biomarker for risk stratification for predicting the prognosis in AMI patients after PCI.The online version contains supplementary material available at 10.1186/s12872-022-02599-5. Acute myocardial infarction (AMI) is the most severe manifestation of coronary artery disease, which causes over 2.4 million deaths in the USA and over 4 million deaths in Europe and northern Asia, and account for over one-third of deaths in developed countries annually . IschemiP = 0.902) [Cystatin C is a potent cysteine protease inhibitor that plays pleiotropic roles in human vascular pathophysiology, in particular regulating cathepsins S and K . Previou= 0.902) .To clarify this question, we conducted a systematic review and meta-analysis to investigate the association between serum cystatin C levels and the prognosis of AMI patients after coronary revascularization.This systematic review was conducted in accordance with the Cochrane Collaboration Handbook, observational studies in epidemiology statement , Meta-AnWe searched PubMed, Embase, and the Cochrane Library until January 21, 2022. Search keywords were “cystatin C,” “coronary artery bypass surgery,” and “percutaneous coronary intervention,” without language restriction. Reference lists of related studies were also manually searched and reviewed to identify other potential studies. This study was completed in accordance with the guidelines of Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) , and theThe included research must meet the following criteria: (1) study included adult patients with AMI who underwent coronary revascularization including PCI or CABG; (2) study reported MACEs and mortality in association with the serum cystatin C level; the MACEs were defined as cardiac death, reinfarction, revascularization, and congestive heart failure; the mortality was defined as all-cause mortality in hospital: any death by fatal cardiac event or other reasons; (3) available data of risk ratio (RR), or hazard ratio (HR) and their 95% confidence intervals (CI) for serum cystatin C level contributed to MACEs or mortality. Exclusion criteria: (1) study with duplicated data, selected the last published article; (2) stable angina pectoris for elective PCI or CABG; (3) conference papers without available data.After excluding duplicates, studies were reviewed to exclude unrelated articles by title and abstract, and then further selected by content. Three reviewers  independently assessed and selected the studies for eligibility. Disagreements were resolved by discussion with the fourth reviewer (LQ). Data were extracted including author, publication year, study type, region, study size, study follow-up, intervention, study outcomes, and adjustment for confounding factors of each study.The Newcastle Ottawa quality assessment scale (NOS)  was usedP > 0.05 or I2 < 50%; and random-effect model for high heterogeneity with P ≤ 0.05 or I2 ≥ 50%). Subgroup analysis analyzed the association between serum cystatin C levels and MACEs according to the intervention (PCI or CABG), ethnicity (Asian vs non-Asian), follow-up duration of study (> 1 year vs ≤ 1 year) and the study sample size (> 500 patients vs ≤ 500 patients). Sensitivity analysis assessed the impact of each study on the results by deleting one study and meta-merging the remaining studies. Sequentially deleting the next study and then meta-merged the remaining ones to test every study’s influence on the final results. If the meta-analysis included more than ten studies, publication bias was assessed by Egger’s test and displayed by a funnel plot. Review Manager version 5.3  and Stata 14.0  were used to conduct our data and perform the meta-analysis.We used the risk ratio (RR) and their associated 95% confidence intervals (Cl) to assess the association between serum cystatin C levels and the MACEs and mortality in AMI patients after coronary revascularization, and P value less than 0.05 was considered to be statistically significant. The effect size conduction was merged by a fixed-effect model or random-effect model according to the heterogeneity of the meta-analysis  and significantly higher mortality risk  in AMI patients after coronary revascularization with statistic difference  from the included studies were pooled together for the association between serum cystatin C levels and MACEs in AMI patients after coronary revascularization in this meta-analysis. The results showed that high serum cystatin C levels were associated with higher MACEs risk  and higher mortality risk  in the Asian AMI population, while, the non-Asian AMI population did not show a significant association with MACEs and mortality, as shown in Additional file Subgroup meta-analysis according to the intervention (PCI or CABG) demonstrated that serum cystatin C was associated with significantly higher MACEs risk  is always accompanied by a serious pathological basis of coronary artery disorder, except for the myocardial infarction with non-obstructive coronary arteries. The alteration of extracellular matrix composition, calcium deposition and advanced glycation end product-mediated collagen cross-linking aggravate the structural arterial stiffness . StructuIn our study, we did not demonstrate a significant association between serum cystatin C levels and the poor prognosis of AMI patients after CABG, which was not consistent with our above hypothesis. However, two included studies about CABG were all showed a significantly higher MACEs risk in the higher serum cystatin C level group. Similar results were observed in the subgroup analysis according to the operation ethnicity (Asian vs non-Asian), Thus, we speculated that it is a statistical dilemma may be caused by limited studies on CABG compared with the studies on PCI. These results would be more accurate and reliable as more CABG studies are conducted in the future.Some limitations should be taken into account in our study. First, the sample size was not large enough among each study, and a total of 7394 AMI subjects were included in the present meta-analysis; however, this is the first and largest meta-analysis for investigating the association between serum cystatin C and the prognosis in AMI patients after coronary revascularization. Nevertheless, we did not further analyse the specific classification  of AMI due to the limited number of studies. Considering that there were some different pathological mechanisms between STEMI and NSTEMI, more studies on different types of AMI are necessary in the future. Second, based on the current limited studies, we have not confirmed a significant relationship between cystatin C and poor prognosis in AMI patients receiving CABG; however, we found that two existing studies have shown that high cystatin C levels are significantly associated with poor prognosis in AMI patients receiving CABG. The results will become clearer as more researches are conducted in the future. Third, sex is another potential risk factor in aging adults, given that older females are reported to be at a greater risk for cardiovascular disease than age-matched men. However, based on current studies, we were unable to perform a subgroup analysis based on sex differences to further explore the effect of cystatin C on the prognosis of AMI patients after coronary revascularization in special group , future research may provide more references based on this topic.Our meta-analysis demonstrated that a higher serum cystatin C level was associated with a higher risk of MACEs and mortality in AMI patients after PCI. Serum cystatin C is a biomarker for risk stratification for predicting the prognosis in AMI patients after PCI.Additional file 1. PRISMA-P  2015 checklist: recommended items addressed in our systematic review and meta-analysis.Additional file 2. Quality evaluation scale for prevalence studies.Additional file 3. Fig. S1: The forest plot of subgroup analysis according to the ethnicity (Asian vs non-Asian) for serum cystatin C contributes for the MACE risk of AMI patients after coronary revascularization.Additional file 4. Fig. S2: The forest plot of subgroup analysis according to the ethnicity (Asian vs non-Asian) for serum cystatin C contributes for the mortality risk of AMI patients after coronary revascularization.Additional file 5. Fig. S3: The forest plot of subgroup analysis according to the follow-up for serum cystatin C contributes for the MACE risk of AMI patients after coronary revascularization.Additional file 6. Fig. S4: The forest plot of subgroup analysis according to the follow-up for serum cystatin C contributes for the mortality risk of AMI patients after coronary revascularization.Additional file 7. Fig. S5: The forest plot of subgroup analysis according to the study size for serum cystatin C contributes for the MACE risk of AMI patients after coronary revascularization.Additional file 8. Fig. S6: The forest plot of subgroup analysis according to the study size for serum cystatin C contributes for the mortality risk of AMI patients after coronary revascularization.