Background: A potential evidence from previous epidemiological studies remains conflicting findings regarding the association between atrial fibrillation (AF) and dementia risk. We, therefore, carried out a meta-analysis of relevant studies to investigate the magnitude of the association between AF and dementia risk.Methods: We performed a systematic literature search of PubMed, EMBASE, and Google Scholar for potential studies between January 1, 1990, and December 31, 2018, with no restriction on the publication language. All potential studies were independently assessed by two reviewers. We only included observational studies that calculated the odds ratio (OR)/hazards ratio (HR) for dementia associated with atrial fibrillation. We first assessed the heterogeneity among study-specific HRs using the Q statistic and I2 statistic. We then used the random-effects model to obtain the overall HR and its 95% CI for all studies. We also tested and corrected for publication bias by funnel plot–based methods. The quality of each study was assessed with the Newcastle Ottawa Scale.Results: A total of 16 studies with 2,415,356 individuals, and approximately 200,653 cases of incidence dementia were included in this study. Patients with AF had a greater risk of incidence dementia than those without AF . Funnel plot and Egger test did not reveal significant publication bias. However, limitations of the study included high heterogeneity and varying degrees of confounder adjustment across individual studies.Conclusion: This study serves as added evidence supporting the hypothesis that AF is associated with an increased risk of dementia. More studies are needed to establish whether optimal treatment of AF can reduce or mitigate the risk of dementia. Atrial fibrillation (AF) is one of the most common cardiac arrhythmia   , diabetes, dietary factors, traumatic brain injury, and multiple environmental factors are associated with incidence and progression of dementia . However, correctly evaluate the extent of the association between AF and dementia risk may have immense clinical inferences for the identification, prevention, and treatment of dementia.CRD42018117343).The protocol for this systematic review was registered in advance with PROSPERO  flow diagram recommended by the Cochrane library  who are expert in meta-analysis subsequently screened all the titles and abstracts for the studies included primarily in our meta-analysis. All the selected studies from this initial screening were then being considered for full-text review. Any disagreement during the screening stage was clarified by discussion with the other experts (HCY and CCW).Studies were considered for inclusion in the meta-analysis if they fulfilled the following criteria:(i). Types of studiesObservational design study such as cohort, case-control, or randomized control trial (RCT).(ii). Types of participantsAn adult (aged ≥ 18 years) with atrial fibrillation (AF).Included studies that reported AF and dementia risk and AF inclusion was confirmed through ECG and physician's diagnosis (ICD-9/10 code), and other standard codes at baseline and each follow-up. Furthermore, for all individuals, it also needed to be confirmed that the presence and onset date of AF was prior to the baseline period and continued during the follow-up periods. A study considered all participants with AF at baseline to have prevalent AF, and also those who developed AF during the follow-up before the diagnosis of any kind dementia.Included studies that reported AF and dementia risk and screened cognitive function using the Mini-Mental State Examination (MMSE) and also provided information regarding the dementia diagnosis based on the DSM-III revised or DSM-IV criteria validated 3-step procedure. We also included studies that examined vascular dementia and AD according to the National Institute of Neurological Disorders and Stroke and Association Internationale pour la Recherch'e et l'Enseignement en Neurosciences (NINDS-AIREN) criteria, and National Institute of Neurological and Communicative Disorders and Stroke and the Alzheimer's Disease and Related Disorders Association (NINCDS-ADRDA) criteria, respectively.Included studies that reported AF and dementia risk and dementia patients was identified by the International Classification of Diseases, Clinical Modification (ICD-9/10-CM) were also included in our study.(iii). Type of interventionAF patient for at least 1 year or longer. Any study with at least a 1 year follow-up period was considered for inclusion. A valid control group was healthy patients without AF and dementia.(iv). Type of outcomePrimary outcome: Development of any kind of dementia in patients with AF.Secondary outcome: Development of AD in patients with AF.We excluded study that did not meet the following criteria: (1) study that was published as a case-report, editorial, review, or clinical trial; and (2) study that did not report dementia risk as their outcome.Data abstraction was conducted by MMI and TNP who used a predefined, standardized guidelines, and data collection procedures. They used the Review Manager software (RevMan-5) to check data accuracy. The following information was garnered from the included studies: (a) method: study design, data collection period, study duration, number of study centers and location, study setting, study protocol, inclusion and exclusion criteria; (b) participants: total number of participants, total number of dementia patients, mean age of participants, age range, gender, percentage of gender, diagnostic criteria; (c) interventions: intervention, comparison, concomitant medications, comorbidities; (d) outcome: primary and secondary findings. Afterward, MMI and TNP garnered the effect size data (HR with 95% CI) from all included observational studies. We resolved disagreements by discussing with the main investigator (YCL).MMI and TNP utilized the Newcastle-Ottawa Scale (NOS) for evaluating the quality of each non-randomized study (cohort study). The NOS scale was used to address the participants selection, study comparability, and outcome or exposure assessment, with scores ranging from 6 to 9 (score 9 considered as no bias) Stang,  Supplem. StudiesA comprehensive subgroup analyses was conducted with type of study design (cohort or RCT), continents (Europe or North America), and methodological quality (good or fair).p < 0.05. Furthermore, the heterogeneity in the results among the studies was calculated using the Higgins I2 which measures the percentage of the total variation across the included studies , version 3, was utilized to conduct statistical analyses. The overall hazard ratios (HRs) from individual studies were pooled using a random-effects model. Furthermore, forest plots were drawn to visually evaluate the results of pooling. AHR value > 1 indicates an increased risk of dementia, HR value 1 indicates no observed association, and HR value < 1 indicates a decreased risk of dementia. In case of statistical significance, Here, Q is Cochran's heterogeneity statistic and df is the degree of freedom.*ES is the weighted effect size that is calculated by multiplying each effect size by the study weight. Negative values of I2 are considered equal to zero. The values of I2 therefore lies between 0 and 100%. A value of 0% indicates no observed heterogeneity. However, a value of I2 at 25–50%, 50–75%, and more than 75% is considered as mild, moderate, and severe heterogeneity  was also calculated to see the extent of variation among the effects observed in different studies (between-study variance). It represents the absolute value of the true variance (heterogeneity)  was conducted using the NOS, which is specifically used to assess the methodological quality of non-randomized studies. We evaluated methodological quality in addressing patients' selection, study comparability, and the assessment of outcome or exposure. We then categorized studies into good, fair, or poor quality. Nine studies received good quality, four studies received fair quality and one study received poor quality . Figure p < 0.0001; I2 = 83.58), in patients with AF. Sixteen studies evaluated the magnitude of the association between AF and dementia risk. The overall adjusted pooled HR of developing dementia was 1.36 . However, lower heterogeneity was observed in this analysis   , 5.n = 10 studies). The overall adjusted pooled HR was 1.39  in the random effect model with higher heterogeneity . However, the overall pooled risk of dementia was 31% for the study with <5 year of follow-up period  in the random effect model with lower heterogeneity .The analyses were also stratified by the study design, region, and follow-up period. While stratified by the length of the follow-up period, the overall pooled HR was appeared to be higher in those studies with more than 5 years of follow-up (p < 0.0001) in the random effect model with higher heterogeneity . The overall risk of dementia of RCT studies was HR 1.31  in the random effect model with lower heterogeneity   .p < 0.0001, I2 = 59.16, Q = 24.48, tau2 = 0.034). The overall risk of dementia of studies (n = 4) published in North America was HR 1.40   .However, the comparison was also stratified by good and fair methodological quality of study; the association of AF with dementia risk was significant in good methodological quality studies, but it appeared to be stronger in fair methodological quality studies .P = 0.90)  .Our meta-analysis provides evidence for a significant association between AF and risk of dementia. This meta-analysis involves a total of 16 unique observational studies with aggregate data on 2,415,356 adult individuals, and approximately 200,653 cases of dementia followed up over a median period of 7.3 years. We found that the presence of AF is associated with an increased risk of dementia (1.36-fold). In addition, when the analysis was stratified either by follow-up duration, study country, or study design, the association between AF and dementia risk appeared to be stronger in studies with a follow-up duration longer than 5 years, and in studies conducted in Asia compared with those conducted in North America and Europe countries. The risk of dementia in AF patients was higher in the cohort studies compared with the one found for the RCT studies. Most of the included studies used MMSE to assess cognitive function, and dementia was clinically diagnosed according to the DSM-IV criteria. Moreover, published studies also diagnosed vascular dementia and AD according to the NINDS-AIREN  criteria and the NINCDS-ADRDA  criteria, respectively.A previous meta-analysis  triggers reduced blood supply in the brain, and also intensifies Aβ production due to dysfunction of the vascular smooth muscle  in the overall main analysis. It is feasible to summarize that high heterogeneity is observed due to differences in sample size, methodological quality, demographic, and ethno-racial characteristics of the study populations, various covariate assessments, different length of follow-up periods, and types of dementia. All these possible sources of statistical heterogeneity were then systematically scrutinized and evaluated using stratified analyses. However, higher heterogeneity was observed between the studies when the primary analysis was conducted but it is notable to mention that there was very low heterogeneity as well as a greater significant association between AF and dementia risk when we stratified our analyses according to study design (RCTs) and different continents (North America and Europe).Our meta-analysis has several limitations that should be mentioned. Although a random-effect model was used in our meta-analysis, the findings of this current and comprehensive meta-analysis should be explained with caution because of their high heterogeneity . Moreover, it is notable to point out that we applied standardized risk estimates from all relevant studies to embrace a harmonious combination of estimates across studies. The greater number of dementia patient's inclusion, stratification according to study design, region, methodological quality, and follow-up duration provided rational statistical power to quantitatively evaluate the association between AF and dementia risk. Finally, some sort of reported bias of the included studies were not a concern in our analyses, as our extensive literature search and discussion with main investigator made it unlikely that any published article was missed, and visual inspection of funnel plots showed no publication bias.The present comprehensive meta-analysis suggests that AF is associated with an increased risk of dementia. It is still unclear whether treatments of AF might lead the greater risk of dementia or not. Therefore, more biological studies are needed to find out their causality.All authors listed have made a substantial, direct and intellectual contribution to the work, and approved it for 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.


P < 0.05). Further Mendelian randomization analysis identified that the amplified risk could be attributed to increased susceptibility (up to 7%) to insulin resistance compared with women. In conclusion, adiponectin measurement might have potential benefits in the detection of metabolic syndrome. Factors that affect insulin resistance should be considered for adjustment in future assessments.Previous studies revealed the potential significance of circulating adiponectin levels with respect to the diagnosis and prediction of metabolic syndrome, but uncertainty has been noted across different cohorts. Systematic evaluation was performed for diagnostic accuracy and predictivity of adiponectin variation for metabolic syndrome in enrolled studies including 1,248 and 6,020 subjects, respectively. Adiponectin can identify metabolic syndrome with moderate accuracy . Heterogeneity analysis revealed that an increasing index of insulin resistance was independently associated with improving the performance of adiponectin upon metabolic syndrome diagnosis . In addition, reductions in adiponectin were associated with increasing metabolic syndrome incidence in a linear dose-response manner. The risk of hypoadiponectinemia with metabolic syndrome was especially increased in men ( Metabolic syndrome (MetS) is defined as a panel of endocrine disorders, including central adiposity, dyslipidemia, hyperglycemia and hypertension Kaur, . Cored bAdiponectin (ADPQ), which is encoded by the AdipoQ gene located on chromosome 3q27.3, is a 244-amino acid pleiotropic hormone with a molecular weight of approximately 30 kDa for its basic subunit. ADPQ is exclusively secreted by adipocytes, and abundant levels in the circulation are eliminated by liver. ADPQ consists of different multimerizations, including low molecular weight (LMW) and high molecular weight (HMW) hexamers. The HMW multimer is considered the major active functional component of ADPQ  accuracy of circulating ADPQ upon MetS diagnosis; (2) causal impacts of varied blood ADPQ on MetS development based on published results. Factors causing heterogeneity were also assessed by subgroup analysis.This study was conducted strictly according to the Preferred Reporting Items for Systematic Reviews and Meta Analyses (PRISMA) criteria , Embase  and ISI Web of Science , were searched up to March 1st, 2018 with the help of an experienced librarian. The criterion for the enrolled studies was that they should be performed with humans and published as original full papers without language restriction.The following subject terms were applied to the literature search: “adiponectin”; “ADIPOQ”; “ADPQ”; “ACDC”; “GBP-28”; “apM1”; “Acrp30”; “metabolic syndrome”; “MetS”; “MS”; “syndrome X” and “insulin resistance syndrome.” More details about the search strategy in a separate database are presented in Table Additionally, relevant papers were searched manually if omitted in routine retrieval but identified in reference lists of enrolled studies.For accuracy analysis, only studies with content that referred to the precision of ADPQ measurement upon MetS diagnosis were considered. In addition, qualified studies should meet the following criteria: 1. necessary data to construct 2 × 2 table for MetS by ADPQ test at best cutoff and largest area under the receiver operating characteristic (AUROC) curves should be provided or extractable by software; 2. MetS was diagnosed with pre-defined criteria.With regard to risk assessment, we only included prospective studies related to the impact of blood ADPQ on MetS occurrence that simultaneously met the following criteria: (1) no patients with MetS at baseline; (2) pre-defined definition for MetS diagnosis; (3) hazard ratios based on ADPQ variation were provided or could be derived from calculations.All studies that met the inclusion criteria were enrolled regardless of study population or sample size. If duplicate data were observed in different studies, we only collected the most recent dataset with detailed information.http://plotdigitizer.sourceforge.net/).Two authors (ZL and SQ) independently extracted information from eligible studies with predesigned unified standardized reporting forms separated by study topics. If relevant information was not provided directly in the enrolled study but presented in attached diagrams, available data would be digitized and extracted via Plot Digitizer software , population characteristics , study design, sampling , definition and prevalence of MetS, and accuracy of ADPQ on MetS diagnosis .Similarly, most items in the accuracy analysis were also included in reported information for risk assessment. However, data accuracy was replaced by MetS risk based on ADPQ variations, including comparison, effect size, calculation and adjusted covariates.The HOMA-IR index was calculated using the following formula when it was not directly provided in the original study: fasting insulin (μU/L) × fasting glucose (nmol/L)/22.5 , specificity (SPE), positive likelihood ratio (PLR), negative likelihood ratio (NLR), diagnostic odds ratio (DOR) and their corresponding 95% confidence intervals (CI) within a random-effect framework based on bivariate original data  extracted from studies  from original studies to evaluate the risk of hypoadiponectinemia associated with MetS occurrence. Quantitative variation of ADPQ by dichotomous covariate  was calculated by standardized mean differences (SMD) and corresponding 95% CI. The overall effect was assessed by integration of MetS risk compared between groups with highest and lowest ADPQ levels. Risk extent was transferred by obtaining the reciprocal of given indicator, i.e., 1/OR as proposed before , sample size, follow-up and disease definition/prevalence was performed on the pooled dose-response risk of ADPQ variation with MetS incidence. The interaction of potential confounders on ADPQ-MetS association was assessed using a Mendelian randomization approach as previously described , whereas a fixed-effect model was used for results with less heterogeneity (I2 < 50%), except for specific illustrations . And quantitative tests of their associated asymmetry were assessed using a regression model  and STATA  software. A two-sided This study was performed and reported strictly according to the statement on Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA)  tool. In general, enrolled studies have high quality for low risk in each item of bias assessment. One study  scale was adopted for quality assessment . All enrThe features of enrolled studies regarding accuracy test and risk assessment are described in Tables Regarding the validity of ADPQ measurements with respect to the diagnosis of MetS, greater than 1248 subjects  were included in the final analysis. Adolescent (mean age range: 11–14 years) and adult participants with various backgrounds  were involved in four and three studies, respectively. Obese participants were enrolled in most (three in four) adolescent studies.n = 6) and prospectively (n = 1) designed. Total ADPQ levels were measured in serum (n = 6) or plasma (n = 1) and assessed by enzyme-linked immunoabsorbent assay  or radioimmunoassay . In all adolescent studies, serum was sampled for ADPQ measurements.The state of insulin resistance was reported or can be evaluated via calculation in a majority of the studies except one  or plasma (n = 2) samples using ELISA (n = 2) or RIA (n = 4) methods. Additionally, two studies , hazard ratio  or both (n = 1) as indicators to estimate the risk measures of MetS occurrence based on ADPQ variation. Risk trends followed with varied ADPQ categories in each individual study are presented in Figure For risk assessment of ADPQ and MetS incidence, six prospective studies including 6,020 general adults were included in the final analysis. Qualified participants included 3,412 men and 2,608 women aged between 31 and 63 years with similar average body mass index (BMI) of approximately 22.8–24.2 kg/mn = 12) are summarized in Table The MetS definitions for all enrolled studies (n = 5) and risk assessment (n = 4). The average ADPQ levels in each MetS group were less than corresponding values from the non-MetS group .The ADPQ value categorized by MetS status was presented in studies related to diagnostic accuracy , PLR (71.6%), and DOR (45.5%) were increased compared with SEN (35.7%) and NLR (39.3%). Different cut-off values across enrolled studies did not influence the test accuracy for an insignificant threshold effect of ADPQ on MetS diagnosis (P = 0.841). Subgroup analysis was performed to investigate the impact of potential confounders on overall effects (Table I2 < 25%). SEN was significantly increased in a cohort of DM patients  with low heterogeneity (I2 < 25%). With respect to SPE, decreased heterogeneity was only observed in the male subgroup (I2 = 0%). Increased SPE was observed in the adolescent cohort . As a joint effect of SEN and SPE, the summary DOR (sDOR) was associated with sample selection, DM status, and HOMA-IR index (P < 0.05). A multi-covariate meta-regression model found that ADPQ measurements in cohorts with higher insulin resistance (HOMA-IR > 5) exhibited better performance in MetS diagnosis .Regarding heterogeneity observed across enrolled studies, the with SEN 5.7% and I2 = 23.6%). In adults, the accuracy of ADPQ in MetS diagnosis was also significant , but unstable effects were noted for mixtures of DM and normal subjects. Inconsistencies in sensitivity and specificity were reduced after excluding the cohort with adult DM patients (both I2 = 0%). Compared with adolescents, significantly reduced AUC of ADPQ for MetS diagnosis was observed in adult cohorts  based on five data points incorporated from four studies  with low inter-subgroup heterogeneity  based on a random-effects meta-regression model corrected by generalized least squares (GLS) estimates (P < 0.01). For data exclusively including men, there was no evidence of a non-linear relationship between circulating ADPQ and MetS . Similarly, the linear trend for the ADPQ-MetS association was also significant in men based on GLS methods in fixed effect meta-regression models (P < 0.01). An approximately 4.6 cases per 1,000 person-year reduction in MetS incidence was associated with a 1 ug/ml increase in ADPQ in male populations .An approximately 2.0 cases per 1,000 person-years reduction in MetS risk was associated with a 1 ug/ml increase ADPQ with moderate inconsistency across individual studies . The reduced OR in male cohorts compared with females partially contributed to the heterogeneity . Intriguingly, a prominent decrease in the dose-response MetS risk was observed in cohorts with lower baseline ADPQ levels that monotonically comprised by men. Further Mendelian randomization analysis revealed the internal differentiation for gender originated from increased susceptibility of IR  in men independent of its specific reduction in ADPQ values . The pooled OR was significantly reduced in two Japanese cohorts . However, the lower MetS risk might result from the lack of BMI adjustment in these studies.Potential confounders of the dose-response MetS risk were investigated by stratification analysis Table . Age, etP = 0.011, Figure P > 0.05).Results from sensitivity analysis also identified that the pooled results exhibited an approximately 3% reduction compared with the primary overall results after omitting the only female study by Kim et al.  (0.91 vsP = 0.379). For risk assessment, convincing evidence indicating significant publication bias across different studies was not revealed based on Egger and Begg's test  despite the slight asymmetry observed in the funnel plot  for MetS occurrence.P < 0.05), possibly due to its distinct sample selection (plasma) and diabetic status of participants upon subgroup analysis. However, heterogeneity tends to originate from DM, but not specimen for high reproducibility and stability of ADPQ as a secretory protein in both serum and plasma  in cohorts with increased insulin resistance (HOMA-IR > 5) also supported this hypothesis regarding insulin resistance as a potential confounder. The influence of insulin resistance on diagnostic value should be noted and adjusted in further clinical practice.Concerning the clear evidence of effects on alleviating insulin resistance  indicates that key factors, such as insulin resistance, should be adjusted. For adults, the efficiency of the ADPQ test upon MetS diagnosis seems inferior for lower AUROC values . The lower diagnostic accuracy might be attributed to the enrollment of normal subjects with a low risk of insulin resistance. Subgroup analysis cannot be performed due to high inconsistency and the lack of a description of insulin resistance in current adult cohorts. More studies in adults with complete baseline features  are required for better risk stratification using ADPQ in MetS diagnosis.ADPQ is closely correlated with MetS in adolescents based on a common influence via the interaction between puberty and adiposity  in men independent of its lower baseline ADPQ level compared with women (SMD: −1.05 ug/ml). Therefore, significant lower OR of MetS incidence associated with ADPQ increased in cohorts with lower ADPQ levels at baseline (< 10 ug/ml, Table Androgens modulate the ADPQ level by inhibiting its secretion from adipocytes to the circulation, and testosterone-related hypoadiponectinemia might contribute to specific severe insulin resistance in men (Nishizawa et al., Centered on insulin resistance, a series of evidence-based studies revealed appreciable variation in ADPQ levels in patients with metabolic disorders. The pooled risks for diabetes and hypertension were approximately 0.72 and 0.94, respectively, with a 1 log transformed/normal unit increase in ADPQ (Li et al., Currently, ADPQ is considered as crucial adipocytokine in MetS development (Okamoto et al., Our study identified ADPQ as a biomarker on detection of MetS and provided evidence about its causative role on the development of MetS. The contribution of ADPQ on the progression and occurrence of MetS can be studied through the use of systems biology tools (Bosley et al., Certain limitations in our study should be noted. First, considering the gender differences, sex-specific diagnostic and predictive validity of ADPQ for MetS should be discussed separately. However, the dose-response OR cannot be pooled given the limited number of prospective studies in women (only one from a Korean population, Kim et al., In conclusion, our study revealed that circulating ADPQ might serve as an available diagnostic biomarker to identify MetS subjects, especially in high-risk populations with insulin resistance. Hypoadiponectinemia predicts increases in MetS incidence in a linear dose-response fashion, and gender should be taken into consideration in this evaluation. ADPQ might be an available target on MetS therapy, but the effects might be differentiated by patient feature.ZL and AM conceived and designed the study. ZL and SQ extracted information. SL analyzed the data. ZL wrote the manuscript. LZ, SZ, and AM reviewed the manuscript. All authors approved the final manuscript for submission.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.


This study is aimed to evaluate the association of KIF6 Trp719Arg variant with angiographic CAD and serum lipid levels in the Han population from northern China. In this case-controlled study, peripheral blood samples were collected from 356 patients and 568 controls of Han Chinese origin. Genotyping was performed by a high-resolution melting curve. The impact of rs20455 on CAD and non-fatal MI was evaluated in a dominant genetic model with stepwise multiple regression analysis. There were no significant differences of genotypes and allele frequency between angiographic CAD and control groups (p > 0.05); however, that of MI and non-MI subgroups were significant differences (p < 0.05). After adjusting for significant risk factors, angiographic CAD risk was not significantly increased in 719Arg allele carriers compared with non-carriers. Further analysis revealed that the non-fatal MI risk and triglyceride levels were significantly higher in 719Arg allele carriers than non-carriers. In conclusion, KIF6 719Arg allele was not an independent risk factor for angiographic CAD susceptibility in Han populations from northern China. However, it was associated with a significantly higher TG level, which may indicate an increased myocardial infarction risk in angiographic CAD patients.KIF6 is a class of molecular motor from the kinesin superfamily. Recently, multiple large studies consisting mainly of Europeans have shown that KIF6 Trp719Arg SNP may be a new predictive factor for coronary artery disease (CAD) event risk. The allelic frequency distribution of rs20455 is different in various populations, yet studies among the Han population, one of the largest ethnic groups in the World, have not been conducted Coronary artery disease (CAD) is influenced by both environmental and genetic factors ,2. In ChFrom 2008 to 2010, there were multiple large prospective and case-control studies on the association of KIF6 Trp719Arg polymorphism with CAD risk. These studies included the cholesterol and recurrent events trial (CARE), including 3,847 Caucasian patients; the West of Scotland coronary prevention study (WOSCOPS) trials , a largeHowever, in recent studies, consisting primarily of Europeans, conflicting results have been reported regarding the association between rs20455 and CAD. Data from a meta-analysis involving 17,000 CAD cases and 39,000 controls from 19 different studies powerfully refuted the rs20455 link to CAD . Also, fThese contradictory conclusions demand evaluation of the predictive ability of this new marker of cardiovascular disease in underrepresented ethnic groups. Importantly, the genotype distribution and allelic frequency of KIF6 Trp719Arg polymorphism has been shown to vary between ethnic groups. The Chinese Han population is one of the World’s largest races, but there have not been related studies on this population. In this case-control study, for the first time, we aim to investigate the KIF6 Trp719Arg polymorphism distribution, its effect on lipid metabolism and association of KIF6 719Arg allele with risk of coronary events  in the Han Nationality from northern China.p < 0.01), smoking status (p < 0.05), diabetes (p < 0.01), HDL-cholesterol (p < 0.05), MI (p < 0.001), angina (p < 0.001) in angiographic CAD cases versus controls.Relevant baseline clinical characteristics and traditional risk factors of the participants are shown in p = 0.982). Among controls, the genotype frequencies were Trp/Trp 29.6%, Arg/Trp 47.2% and Arg/Arg 23.2%; whereas in angiographic CAD patients, the genotype frequencies were 29.2%, 46.1% and 24.7%, respectively. There were no statistically significant differences in genotype frequencies between angiographic CAD cases and the control group (p > 0.05). The frequency of 719Arg was not significantly different between angiographic CAD patients (47.8%) and controls (46.8%) .924 participants were successfully genotyped for the KIF6 Trp719Arg polymorphism . As showp < 0.05). The frequency of 719Arg was significantly different in MI patients (56.1%) compared with non-MI patients (43.8%) . Also, there were significant differences in the minor allele frequency between MI (56.1%) and Non-MI + control subgroups (45.9%) .The angiographic CAD patients were further divided into MI and non-MI subgroups. As shown in p = 0.004), smoking status  and diabetes  were major independent risk factors for angiographic CAD in these populations, while the plasma HDL-cholesterol levels were a protective factor .To identify traditional risk factors for angiographic CAD, stepwise multiple regression analysis was performed, which included the variables shown in p = 0.9203) compared with non-carriers.As shown in p < 0.0001) in carriers of the KIF6 719Arg allele, compared with non-carriers between MI and non-MI subgroups. On comparison of the MI and non-MI + control, statistically significant differences in MI risk were also evident .As shown in p < 0.05). In controls, LDL-cholesterol levels in carriers of 719Arg variant were significantly higher when compared to non-carriers (p < 0.05). In contrast, no significant difference was observed in TC or HDL-cholesterol levels among genotypes in all groups (p > 0.05).As shown in p > 0.05). When evaluating the difference in genotype between control and angiographic CAD groups, we failed to find a significant difference in genotype distribution and allele frequencies of KIF6 Trp719Arg SNP. Binary regression analysis also revealed that 719Arg allele may not be an independent risk factor for angiographic CAD in the Han people of northern China. In a subgroup analysis restricted to the angiographic CAD group, on the effect of 719Arg allele on non-fatal MI risk, the genotype distribution and minor allele frequency showed significant differences. We also found that the susceptibility of non-fatal MI was increased significantly compared with non-carriers independent of various risk factors such as age, sex, and smoking status. This is the first report on the association of KIF6 Trp719Arg polymorphism with angiographic CAD events in Han populations from northern China. Nine hundred and twenty four (924) participants were successfully genotyped for the KIF6 Trp719Arg polymorphism. To our knowledge, this is the first publication describing HRM analysis as a detection method for the rs20455 polymorphism. As a new detection method, HRM analysis is associated with several advantages over traditional genotype analysis, including faster speed, simple operation, high throughput, high sensitivity, high specificity, and limited pollution. In our case-control study, the minor allelic frequency of rs20455 is 0.478, which is lower than Sub-Saharan African populations (0.908) or Han Chinese (0.556), but it is higher than in European (0.358) and Japanese (0.386) populations . The genet al. in 1985 [Kinesin is a molecular motor that was discovered in squid and mammalian nervous tissue by Vale  in 1985 ,22,23. T in 1985 . Trp719A in 1985 ,13,20,25Results from several clinical studies have also contradicted the association between Trp719Arg and CAD . In 2010p > 0.05) [CAD is the second leading cause of cardiovascular death in China . Accordi > 0.05) . To dateGenetic polymorphisms associated with CAD events are primarily related to lipid metabolism of lipoproteins, apolipoproteins and enzymes involved in gene encoding. To assess whether an increased risk of MI/angina in carriers within the CAD population was associated with abnormal plasma lipid level, we performed analyses of the plasma lipid levels between genotypes. The recently published regression meta-analysis of 144,931 participants that included KIF6 negative studies , reported that the 719Arg allele increased vulnerability to the harmful effect of LDL cholesterol on the risk of CVD . Our resThere were several limitations of this study. The small number of patients and unequal baseline data may have confounding effects. A larger sample size and comprehensive analysis of the interaction between genetic and environment factors may reveal additional insight into the role of rs20455 in the pathogenesis of CAD. Additionally, further study should be designed to test the association of KIF6 variants with statin response in Han Chinese.The present case-control study included a cohort of 924 unrelated Han Chinese residents recruited from the Changchun, Jinlin area in Northern China. The cohort was comprised of 356 CAD subjects with the following inclusion criteria: coronary angiography of >50% stenosis in one or more arteries and stable or unstable angina. MI was defined by detection of elevated cardiac biomarkers (upper reference limit for creatine kinase-MB is 0–25 U/L or troponin is 0.1 ng/mL) together with evidence of myocardial ischemia . Five huBaseline clinical information was obtained by reviewing the subject’s medical records. Patients with diabetes mellitus were identified as having fasting plasma glucose >6.9 mmol/L, or using anti-diabetic medication. Hypertension was deﬁned as systolic blood pressure >140 mmHg, a diastolic blood pressure >90 mmHg, or patients with a documented diagnosis or using antihypertensive medication. All laboratory parameters were determined in fasting patients not using lipid-lowering medication for at least four weeks. All lipid levels were measured on Hitachi 7170 automatic biochemical analyzer. Total cholesterol, triglyceride were measured by enzyme method, reagent by DESAY diagnostic system  Limited company. LDL-C and HDL-C levels were detected by direct determination method, using reagents provided by Beijing Jiu Qiang Biological Limited .2 (25 mM) 1 μL, dNTPs (2.5 mmol/L) 0.25 uL, Eva-green saturation dye (20×) 0.5 μL, primer (10 μmol/L) 0.25 μL, template DNA (10 ng/uL) 1 μL, Taq enzyme (5U) 0.1 μL, ultra pure water 5.65 μL. The expansion program was 95 °C for 5 min (pre-degenerated), 95 °C for 10 s, 60 °C for 15 s, 72 °C for 25 s (centigrade), for 50 cycles. Genotyping was performed by a high-resolution melting curve according to the Roche Light Cycler 480 protocol: 95 °C for 1 min, 40 °C for 1 min, 95 °C for detecting fluorescent 40 times per second, cool to 40 °C for 10 s.Blood samples were drawn by vein-puncture after an overnight fast of 12 h. The samples were separated by centrifugation, and aliquots were stored at −86 °C until analyses. Genomic DNA was extracted following a standard protocol . Primerst-tests (for continuous variables), by chi-square statistics or Fisher’s exact test , and reported as two-sided p-values. Two-sided p-values < 0.05 were considered statistically significant. Deviations from Hardy-Weinberg Equilibrium expectations were tested by the chi-square test. The gene counting method tested genotype distribution and allele frequencies [versus non-carriers (Trp/Trp).All data were analyzed using SPSS 17.0 . Differences between baseline characteristics of participants were assessed by quencies . The chiIn conclusion, our results supported the notion that carriers of the KIF6 719Arg allele are not associated with risk of angiographic CAD, but have an intrinsic correlation with an increased risk of non-fatal MI in angiographic CAD subjects of Han nationality from China. Carriers of the 719Arg allele have higher TG levels than non-carriers, indicating that KIF6 Trp719Arg SNP may be involved in lipid metabolism and may serve as a MI diagnostic biomarker in Northern Chinese populations. The functional research regarding the KIF6 variant with respect to angiographic CAD, and the benefit from statin therapy, awaits confirmation in future studies. Overall, these results may indicate that rs20455 could predict those at highest risk for MI, which may be important in pre-clinical diagnosis, early risk stratiﬁcation, early intervention, prognostication, or individualization of cardiovascular therapy.


NOS3) G894T polymorphisms, air pollution and their interaction on ischemic heart disease (IHD) risk across populations worldwide. Recursive partition trees, nonlinear association curve fit and geographic information system maps were incorporated to verify results of conventional pooled analyses for sources of heterogeneity. Results from 61 studies  revealed a significant increased relative risk (RR) of IHD associated with NOS3 894 polymorphisms TT (RR = 1.44) and GT (RR = 1.37). Subgroup analysis revealed that the TT polymorphism genotype had significantly increased risk of IHD in Caucasian, East Asian, South Asian, and Middle Eastern populations . It is important to point out that many countries demonstrated an average risk of greater than two, which identifies the NOS3 894 TT polymorphism as a potential causal factor and biological marker of IHD, based on criteria for strong evidence used in international consensus panels. These 10 countries include Ukraine, the United Kingdom, Brazil, Chile, Japan, South Korea, India, Iran, Egypt and Morocco. For these countries with elevated risk (RR > 2) from the NOS3 894 TT polymorphism, meta-predictive analysis demonstrated an increasing trend in air pollution association with increased NOS3 894 polymorphisms. Further studies are needed to explore the complexity of the associations among NOS3 gene polymorphisms per population stratifications within countries, detailed air pollution data for added specificity for geographic location across time, and disease risk.The purpose of this updated meta-analysis was to investigate the effect of nitric oxide synthase-3 ( Ischemic heart disease (IHD) is the leading cause of death and a major economic burden worldwide. It is the cause of over 30% of total annual deaths ,2,3,4 anNOS3 894GT located in exon 7  is a genetic marker that has been specifically linked to an increased risk of IHD, hypertension, coronary spasms, and stent re-stenosis [NOS3 894GT SNP represents a guanine (G)/thymine (T) substitution at position 894 on exon 7 leading to a change from glutamate to aspartate at position 298; rs1799983 [NOS3 G894T polymorphisms [NOS3 G894T with the development of IHD among various ethnicities and disease subgroups [Endothelial-derived nitric oxide (NO) is a powerful vasodilator with known vasoprotective effects, and its physiologic functions are as follows: (1) vascular relaxation; (2) platelet suppression; (3) decreased leucocyte adhesion to the vascular endothelium; (4) reduced smooth muscle cell proliferation and migration ,10,11,12stenosis ,15. The orphisms , with prubgroups ,19,20,21NOS3 [NOS3 affects metabolism in the urea cycle of the methylation pathway, which is critical for preventing systemic inflammation as an epigenetic risk factor for heart health, it is important to explore associations among NOS3 894 gene polymorphisms, air pollution, and IHD [Likewise, there are several epidemiologic studies suggesting the incidence of IHD morbidity and mortality may be related to exposure to ambient pollutants that trigger the development of IHD ,27,28,29NOS3 . Because and IHD ,45,46,47NOS3 894 polymorphisms as risk factors for IHD, using meta-predictive analytics to investigate the source of heterogeneity, including air pollution for IHD susceptibility. This study is the most comprehensive meta-analysis to date on the impact of NOS3 894 on IHD. The gaps in knowledge were filled that were not addressed in previously published meta-analyses which focused only on sub-populations (Asians) and sub-types of IHD.Therefore, the purpose of this updated meta-analysis is to examine the association of NOS3 894, ischemic heart disease, coronary heart disease, meta-analysis, case-control, and was limited to human studies. In addition, previous meta-analyses and review papers were used to cross reference and trace back to all original studies. Databases were searched repeatedly every three months for a total of four different times to identify any new papers. Two investigators were responsible for conducting the literature search and recording data. Studies published in other languages or from other countries were included, but only if they included both an abstract in English and tables clearly listing genotype allele counts. Studies lacking genotypes for case and controls or appropriate genotype allele counts were excluded.Online databases of PubMed, PubMed Central and Google Scholar were searched for all available studies from 1998 (publication year of first related study) through September 2017 following the guidelines for preferred reporting of items for meta-analysis of observational studies ,49,50. TTwo hundred and forty-two relevant studies were initially identified, with 106 subsequently excluded because they did not have genotypes for case and controls and/or genotype allele counts. The 136 remaining studies were retrieved for evaluation. Of those, 34 studies were further excluded due to no English abstract and/or no appropriate genotype allele counts. The remaining 102 studies were retrieved for further evaluation, of which 49 studies were further eliminated due to no clear genotype counts. Further searches resulted in the identification of eight additional usable studies from a list of 30 studies. Thus, a total of 61 studies were included in the final meta-analysis .Study populations were drawn from Australia, Europe, Central America, South America, Asia, the Middle East, and Africa. The majority of populations from these studies were Caucasian (25) and East Asian (18) studies, followed by South Asian (six), African (six), Middle Eastern (four), and Hispanic (two) . The airNOS3 894 genotypes are reported in Data including genotype allele counts and quality scores were checked for accuracy by three team members; for discrepancies, a consensus was reached by involving a fourth team member based on the criteria for quality in published meta-analyses (PRISMA) and other observational studies ,49,50. TNOS3 894 genotypes for cases and controls and 95% confidence intervals (CI) were calculated for the associations of polymorphisms and IHD with heterogeneity tested before selection of an association model. Pooled RRs are preferred to odds ratios (OR) according to related consensus reports for the field of study [NOS3 894 genotypes  as the denominators to compute standardized ratios for RRs. Thus, a pooled RR of greater than one represents an increased risk for IHD among gene variations, whereas a pooled RR less than one indicates a protective effect (favoring the case or the IHD group). Significant findings were identified at a value of p < 0.05.Genotype allele counts, along with all other data, were entered into Excel , and pooled using Stats Direct Version 3.1.4 . Pooled risk ratios (RR) for of study ,59,60; tof study ,62,63. WNOS3 894 polymorphism rates and IHD risk; GIS maps were generated to manage the geospatial data set and help associate regional patterns of polymorphisms and IHD risk with AP death by country [NOS3 894 polymorphism rates in cases and control groups and risks) by splitting data into subgroups based on levels of AP death [Additionally, because the data presented heterogeneity with some regional differences on polymorphism rates and risks, the JMP 13 Pro program  was used to generate geographic information system (GIS) maps to depict global gene polymorphism patterns and IHD risks, allowing visual identification of geographic patterns . JMP is  country ,64. RecuAP death ,33,66. WAP death . In otheAP death , and thu2 tends to yield a complex and over-fitted model because R2 always goes up with additional predictors. On the other hand, AIC or Akaike’s information criterion correction (AICc) does not necessarily change with the addition of variables. Rather, it varies based upon the composition of the predictors; thus, it is more likely to yield an optimal model [To provide consistency with other meta-analyses and allow for easy comparison, we also employed a conventional multiple comparison procedure (Tukey’s test)  to examial model .NOS3 genotypes from 61 studies with genotypes for case and controls was used to identify the association of NOS3 894 gene variations with IHD in global populations. The analysis included a total of 16,219 IHD cases and 12,222 controls (p < 0.0001). The heterozygous GT polymorphism genotype (37.4% cases and 32.4% controls) was associated with an increased risk of IHD for all groups . The homozygous TT polymorphism genotype (9.5% cases and 6.2% controls) presented the least frequently, and, like the GT polymorphism, was associated with an increased risk of IHD for all groups . Furthermore, the T allele was associated with an increased risk of IHD for all groups combined . In contrast, the G allele was found to be protective against IHD for all racial-ethnic groups . Homozygous TT polymorphism percent distributions were observed to be higher in Caucasians (10.2%) than in African (8.0%), Hispanic (4.4%), Middle Eastern (4.2%), South Asian (2.0%) and East Asian groups (0.7%), in the control groups as the basis for healthy populations. GT polymorphism percent distributions were observed to be higher in Caucasians (43.6%) than in African (38.2%), Hispanic (32.8%), Middle Eastern (30.6%), South Asian (24.4%) and East Asian groups (14.6%), in the control groups as the basis for healthy populations. Similar distinctive patterns were noted for the IHD cases across all of the racial-ethnic groups. NOS3 894 TT and GT genotypes as risk genotypes (RR > 1) of IHD. Racial-ethnic subgroup analysis revealed that the homozygous TT genotype was associated with an increased risk of IHD in Caucasian (p = 0.0051), East Asian (p < 0.0001), South Asian (p = 0.0076), and Middle Eastern (p = 0.0003) populations. The heterozygous GT genotype was also associated with an increased risk of IHD for both East Asian (p < 0.0001) and South Asian (p = 0.0304) populations.Pooled analysis of controls . The GG NOS3 894 TT genotype split by countries that demonstrated a risk type  and countries with a protective type (RR < 1)  . Subgrou(RR < 1) ; nine co(RR < 1) ; and fou(RR < 1) . When vir of IHD ,60,61,62 Tunisia .NOS3 894 polymorphism rates and IHD risk association and enabled visual detection of regional patterns. The GIS maps graphically depict the highest total percent of NOS3 894 TT and GT genotypes for both control and cases groups in Caucasian nations  was used to select the best model with good fit to the truth with simplicity and parsimony [NOS3 894 polymorphisms and heart disease. Nonlinear association methods were used to explore the association between AP death and the NOS3 percentage of polymorphisms  [MTHFR gene is known to be heat sensitive [NOS3 mechanism; however, the trend of increased NOS3 polymorphisms with increased air pollution for countries with higher risk (RR > 2) is consistent with each of these previous studies [NOS3 polymorphisms; however, Asian countries presented with higher IHD risk in association with NOS3 894 polymorphisms.Our meta-predictive analysis demonstrated an upward trend in air pollution death rate level associated with the ase risk ,45,46,47cy (HDP) ,33,34,35ensitive ,85, ther studies ,33,34,35NOS3 894 TT and GT , whereas the common allele NOS3 GG genotype was observed to be protective for all groups. The T allele was associated with an increased risk of IHD for all groups, and, consistent with reports from previous meta-analyses, suggests those carrying the NOS3 894 T allele may have a predisposed risk for IHD [NOS3 894 TT polymorphism genotype as a potentially causal factor of IHD as a biological marker based on the criteria for strong evidence used in international consensus panels [NOS3 894 polymorphisms.Our analysis revealed significant associations between  for IHD ,19,20,21s panels ,60. For NOS3 894 polymorphism and MI risk among Asians vs. non-Asians [NOS3 894 polymorphisms on IHD risks could undergo several potential pathways  in Asian populations [NOS3 894 polymorphisms and two leading risk factors of IHD, hypertension and type 2 diabetes in Asians [NOS3 894 on IHD risk among various ethnic groups/regions.Heterogeneity with regional differences on polymorphism rates and IHD risk was demonstrated by the data. Congruent with our study, others have also identified ethnicity as the source of heterogeneity in the association between n-Asians . Mechaniulations ,87. Thern Asians ,89,90,91n Asians ,93. AddiNOS3 894 polymorphisms and associated IHD risk across global populations. This analysis revealed a trend in the association between increased air pollution associated death rates with increased NOS3 894 polymorphisms. These findings provide novel insight regarding the association between NOS3 894 polymorphisms, air pollution, and IHD risk. Both homozygous TT and heterozygous GT genotypes of NOS3 894 in Caucasian populations were found to be higher than in other populations including African, Hispanic, Middle Eastern, South Asian and East Asian. However, both homozygous and heterozygous NOS3 894 polymorphism-associated IHD risks are higher in Asian nations when compared to other nations. Future studies exploring epigenetic factors in association with NOS3 894 polymorphisms and IHD and prevention strategies for IHD are needed.The results of our comprehensive meta-predictive analysis demonstrated differences in the rate of


Previous studies have suggested that metabolic syndrome (MetS) and its component conditions are linked to the development of many benign or malignant diseases. Some studies have described relationships among metabolic syndrome or diabetes and liver cancer, but not many articles described the relationships between MetS and cirrhosis, acute hepatic failure, end-stage liver disease, and even death. However, liver cancers, cirrhosis, acute hepatic failure, end-stage liver disease, and liver-related mortality—collectively described as liver-related events (LREs)—may have different relationships with MetS. We undertook this meta-analysis to examine the association between MetS and LREs, and to determine whether geographic region or hepatitis B virus (HBV) positivity might influence the association.Relevant studies were identified from PubMed, EMBASE, and the Cochrane database. Two reviewers independently searched records from January 1980 to December 2017. The search terms included ‘metabolic syndrome’, ‘diabetes mellitus’, ‘insulin resistance syndrome’, and ‘metabolic abnormalities’, combined with ‘cirrhosis’, ‘hepatic fibrosis ’, ‘hepatocellular carcinoma’, ‘complication’, ‘LRE’, ‘HCC’, ‘liver-related events’, and ‘liver cancer’. No language restriction was applied to the search. We chose the studies reporting an association between MetS and LREs. We used Begg’s and Egger’s tests and visually examined a funnel plot to assess publication bias. All analyses were conducted in Stata 14.0 software.p values showed a publication bias in this meta analysis. However, through the trim and fill method, we obtained a new RR value for LREs with MetS of 1.49 . There was no obvious difference with the two answers, so we concluded that the results were robust. For hepatitis B positive patients, the RR for MetS and LREs was 2.15 , but for the hepatitis B negative patients, the RR was 1.85 . And for non-Asians, the RR for MetS and LREs was 2.21 , while for Asians, the RR was 1.73 .There were 19 studies (18 cohort and 1 case-control) included in the analysis, with a total of 1,561,457 participants. The subjects’ ages ranged from 18 to 84 years. The combined analysis showed an overall 86% increase risk of LREs in cases with MetS . The funnel plot was asymmetrical, and the Egger’s test This meta-analysis showed that MetS is associated with a moderately increased risk of LREs prevalence. Patients with MetS together with hepatitis B are more likely to develop hepatic events. For non-Asians, MetS is more likely to increase the incidence of LREs. Liver cancer is the most severe liver disease. Although MetS is known to promote liver cancer, there is little evidence of whether MetS is associated with cirrhosis, liver failure, liver fibrosis, and death from liver causes.Because of a significant increase in the incidence and mortality of hepatocellular carcinoma (HCC), this cancer has become one of the most common malignancies and a major cause of death worldwide. In recent years, chronic liver disease has become a major cause of death in the United States, causing a large number of deaths every year, according to national life statistics. All the cirrhotic complications, HCC, and/or liver-related mortality were called LREs. And liver-related death is defined as death related to hepatic events . Viral hMetS has become a major public health focus worldwide, and it is related to the occurrence of obesity and the diabetes pandemic. MetS components are a series of risk factors for cardiovascular diseases and it has become an increasingly severe problem globally . The risIn this study, we conducted a systematic review and meta-analysis to assess all available evidence to identify an association between MetS and LREs. We additionally analyzed related factors.We used PubMed, EMBASE, and the Cochrane database for literature retrieval. In December 2017, the following search terms were used without language restrictions: MetS, diabetes mellitus, insulin resistance, metabolic abnormalities, cirrhosis, hepatic fibrosis, hepatocellular carcinoma, complication, LRE, HCC, liver-related event, and liver cancer All references cited in these studies were also reviewed to identify other published articles not indexed in the databases. The systematic review process followed established quality standards for reporting of meta-analyses.The inclusion criteria for studies in the meta-analysis were as follows: either case-control or cohort design; inclusion of subjects over 18 years old; assessment of the effects of MetS on the risk of liver events; and reporting of relative risk (RR) estimates for LREs in subjects with MetS. Only complete papers and published studies in the medical literature were included. Data from summaries, reviews, editorials, case reports, and letters were excluded. Studies reporting no risk ratio (HR) and 95% confidence intervals (95% CI), and those with participants with incomplete data were also excluded.The following data were collected from each study: first author’s surname, type of article, year and country of publication, distribution of age and sex, number and characteristics of the participator, definitions of MetS, risk estimates with corresponding confidence intervals (CIs), and all other information. Quality assessment for cohort studies in this meta-analysis was assessed with the Newcastle Ottawa scale (NOS), as recommended by the Cochrane Non-Randomized Studies Methods Working Group. The scale allocates a number of stars ranging from one to nine.I2 statistic and p value. Funnel plots and Egger and Begg’s and Mazumdar’s tests were used to assess publication bias. A p value < 0.05 was considered statistically significant [We used both fixed- and random-effects models to calculate the pooled RR and 95% CI. We performed an analysis to identify the association between MetS and LREs, and also to analyze any influence of geographic region or HBV positivity. Heterogeneity was assessed with the nificant , 8. The Figure A total of 19 articles were included in this meta analysis. Among the 18 cohort studies, there were 1,546,392 participants, of which 7373 had liver-related events (LREs)    Fig. .Fig. 4Fop = 0.0005, I2 = 86.2%)  Fig. .Fig. 6FoWe divided all the data into Asian and non-Asian regions, and then calculated the relationship between MetS and LREs in each group. We then compared the differences between the groups.p = 0.000, I2 = 76.0%)  Fig.  .Fig. 8Fp values of Egger’s test were close to 0.05 . Thus, the results suggested that publication bias was present in this meta analysis. Then we use the trim and fill method to correct the funnel asymmetry caused by publication bias, and we get a new RR for MetS and LREs which is 1.49, and there was no obvious difference compared with our previous results.We used the NOS to assess publication quality and determined credible results with 8.2 stars for the 18 cohort studies and 8 stars for the single case-control study Table  and EggeIn the meta-analysis of 19 studies including 1,561,457 patients with MetS, 16 studies found a statistically significant positive association, while 3 studies showed a negative association. After we removed the negative study indicated by the random-effects model to have affected the results, we also found that MetS significantly increased the risk of LREs. Therefore, the negative study did not change the finding.p values of Egger’s test showed a publication bias in this meta analysis. However, through the trim and fill method, we obtained a new RR for LREs with MetS of 1.49 , and there was no obvious difference compared with our previous results. Therefore, we concluded that the results were robust, and MetS is associated with a moderately increased risk of LRE prevalence.The funnel plot was asymmetrical, and the The results also indicated that preexisting MetS confers a statistically significant 1.56- or 2.23-fold increased risk for LREs that is independent of other risk factors, an association also observed for its components. MetS can promote the occurrence of cardiovascular diseases, coronary heart disease and other diseases, and the degree of association between MetS and liver cancer is similar to the former , 27. NonThe findings from our study suggested a 1.33- or 2.33-fold increased risk of HCC development in patients with MetS. In addition, the study also demonstrated a 1.55- or 3.74-fold increased risk of liver-related death in patients with MetS. Our results together indicate that MetS is associated with a moderately increased risk of LRE.Analysis of related factors showed an overall 115% increase in risk of LREs in HBV-positive cases . For HBV-negative cases, the probability indicated a 85% increase in risk . This conclusion is consistent with viral hepatitis having been identified as a major risk factors for LREs [And there was some article showed the pooled RR for HCV subjects with IR was 1.89 . It is because the presence of the HCV core protein will increase the level of tumor necrosis factor-α, further leading to proteasomal degradation of the insulin receptor substrate, ultimately altering insulin function and promoting the development of IR. At the same time, IR further causes lipid accumulation in the liver and production of reactive oxygen species, and indirectly activates stellate cells, eventually leading to the occurrence of liver fibrosis .The related factor analysis also indicated that non-Asian MetS populations are more likely than Asian populations to have LREs. There was a 73% increase in risk of LREs for Asians and an 112% increase for non-Asians. The reason for this result is not yet clear, and more research is needed to provide an explanation. It is possible that the economy and the typical non-Western diet in Asia may explain the lower risk of MetS. And for Asian, the HBV vaccine is being recommended widespread. Meanwhile, the incidence of T2DM is lower in the Asian population, and it may have influenced the results. Some studies have reported that up to 90% of obese people in Europe have some degree of fatty degeneration in their liver and overweight increases the risk of HCC. The observed association between excess body weight and the increased risk of liver disease seems to support the reports that liver disease in obese individuals may be mediated through the development of NAFLD and NASH [Our study suggests a relationship between MetS and LREs, but the mechanism that links the two is not fully understood. MetS is likely to be a proxy for other cancer risk factors, such as low physical activity, intake of high caloric food producing a high quantity of heat, high dietary fat intake, low fiber intake, and oxidative stress .The mechMoreover, diabetic people have a higher risk of HCV infection, and diabetes is also associated with hyperinsulinemia. Both MetS and LREs are a problem worldwide, and growing evidence shows a relationship between MetS and an increased risk of LREs . InsulinA previous meta-analysis has described a relationship between hypertension and cancer, thus suggesting that hypertension is associated with an increased risk of cancer death . That stThere are several limitations that should be considered when interpreting the findings of our meta-analysis. First, there are multiple factors that were not considered in the combined RR analysis, such as dietary habits, HCC family history, and other genetic risk factors, owing to the unavailability of these variables in the original study. Second, for diabetes and liver cancer risk meta-analysis, we referred data from observational studies, while those studies may have lower heterogeneity. Third, because chronic liver disease can also cause diabetes, the link between diabetes and HCC may not be exact. Fourth, different controls on confounding factors also influenced the results. Finally, the meta-analysis was limited to English-language studies, which might have introduced publication bias.However, our study has several strengths; for example, this article included more data than previous studies and examined links between MetS and cirrhosis, as well as liver-related deaths, among other factors. We also analyzed the geographic region and the population for hepatitis B and other parameters. The all cohort studies in our meta-analysis are of high quality and were able to detect potential associations and eliminate the possibility of recall and selection bias.In conclusion, this population-based study showed that MetS is an important risk factor for the development of liver disease. Therefore, the control of the globally prevalent MetS may help reduce liver disease. At the same time, the metabolic syndrome and related factors are introduced in the paper. Trying to avoid these risk factors is a better treatment. More well-designed randomized controlled trials or prospective cohort or retrospective cohort studies are urgently needed to improve understanding of this risk.


Many studies have tried to elucidate the connection between vitamin D receptor (VDR) gene (ApaI) polymorphism and periodontitis; however, so far there is no consensus. To further assess the impact of ApaI polymorphism on periodontitis risk, we have conducted a meta‐analysis of Chinese population. Relevant literatures were searched according to PubMed and Chinese database in January 2019. The strength of correlation was evaluated by combining odds ratio (ORs) and 95% confidence interval (CIs). Six case–control studies were identified with inclusion criteria, including 734 cases of periodontitis and 687 controls. Based on the overall analysis, the VDR ApaI polymorphism was not due to the risk of periodontitis in all models. Subgroup analysis showed that the risk of periodontitis in North China was significantly reduced. To sum up, the study shows that VDR‐ApaI polymorphism may be connected with a lower risk of periodontitis in northern China. It is suggested that inferential studies should be conducted in other ethnic groups. Searches are conducted without language restrictions, with emphasis on human research. References cited in the retrieved articles are also included.The criteria for inclusion are as follows: (a) case–control or cohort studies describing the correlation between VDR ApaI polymorphism and periodontitis; (b) genotype data of the VDR ApaI polymorphism in periodontitis patients and controls; and (c) participants are Chinese. The criteria for exclusion are as follows: (a) redundant literature; (b) absence of controls; (c) incomplete data; and (d) meta‐analyses, letters, meeting abstracts, reviews, and editorial articles.2.2Two reviewers extracted data from each study. The dispute was stable through discussion. Screen the headlines and abstracts of all potentially relevant manuscripts to determine relevance. If the headline and abstract are ambiguous, check the whole article carefully. We collected the information as follows from each study: first author's surname; publication year; type of periodontitis; source of control; geographic region (s); sample capacity; and quantity of subjects with VDR ApaI genotype.2.3Z test was applied to determine the significance of combined ORs and 95% CIs. The heterogeneity was measured by Q statistics. In cases of heterogeneity, the random‐effect model was selected to aggregate 95% CIs‐distributed ORs, and then, the fixed‐effect model was applied.The statistics of correlation were evaluated by combining odds ratio (ORs) and 95% confidence interval (CIs). p < .05 was considered as the statistical significance.Sensitivity analysis was evaluated by contrasting the results of the fixed‐ and random‐effect models. In addition to comparing all subjects, stratified analysis was conducted according to geographic region(s), control source and type of periodontitis. Statistical analyses were handled by the University City of Texas Stata Corporation . 33.1Totally 62 studies that examined the relationship between the VDR polymorphisms and the risk of periodontitis were distinguished after document duplication was deleted in different databases Figure . Fifty a3.2The primary results are enumerated Table . Firstly3.3Statistics analysis was conducted by comparing the results of fixed‐effect and random‐effect models. Except for the allele control model, all significant corresponding ORS remained unchanged, indicating that the outcomes of the meta‐analysis were relatively steady Table .4p < .001)  on periodontitis and VDR ApaI risk; however, several shortcomings should be taken notes. First of all, ethnic‐specific meta‐analysis only includes figures on a single ethnic group, so our study applies only to that ethnic group. Secondly, because this meta‐analysis is mainly based on unadjusted impact estimates and cis‐index, the interferences are uncontrolled. Thirdly, the heterogeneity is high and stratified analysis cannot explain it. Other clinical heterogeneities may help to diagnose and classify periodontal diseases, differences in periodontal examinations, and different clinicians; however, due to the limited data available, we cannot investigate these factors. Finally, owing to the limitations of funnel charts, which require a large series of studies, we could not assess publication bias in this meta‐analysis.All in all, the meta‐analysis suggests that VDR ApaI polymorphism may be connected with a reduced cause of periodontitis in northern China; however, much more researches should be investigated in the future to verify our findings.The authors declare that they do not have any conflict of interest.This study does not involve any human or animal testing, and all analyses are based on previously published studies; thus, no ethical review and patient consent are required.


The effect of sleep duration on cancer risk remains controversial. We aimed to quantify the available evidence on this relationship using categorical and dose–response meta-analyses.Population-based cohort studies and case-control studies with at least three categories of sleep duration were identified by searching PubMed, EMBASE, and the Cochrane Library database up to July 2017.Sixty-five studies from 25 articles were included, involving 1,550,524 participants and 86,201 cancer cases. The categorical meta-analysis revealed that neither short nor long sleep duration was associated with increased cancer risk . Subgroup analysis revealed that short sleep duration was associated with cancer risk among Asians  and long sleep duration significantly increased the risk of colorectal cancer . The dose–response meta-analysis showed no significant relationship between sleep duration and cancer risk. When treated as two linear piecewise functions with a cut point of 7 h, similar nonsignificant associations were found .Categorical meta-analysis indicated that short sleep duration increased cancer risk in Asians and long sleep duration increased the risk of colorectal cancer, but these findings were not consistent in the dose–response meta-analysis. Long-term randomized controlled trials and well-designed prospective studies are needed to establish causality and to elucidate the mechanism underlying the association between sleep duration and cancer risk.The online version of this article (10.1186/s12885-018-5025-y) contains supplementary material, which is available to authorized users. Cancer is a major public health problem worldwide and a leading cause of death in both more and less economically developed countries; the global burden of cancer is expected to increase because of population growth and aging . Some liSleep is an essential component of healthy development and necessary for physical and mental health . IncreasThe objective of this meta-analysis was to update the evidence on the relationship between sleep duration and cancer risk. We also aimed to explore the quantitative estimates, determine the overall shape of the relationships between sleep duration and cancer incidence, and compare categorical and dose–response meta-analyses.The systematic review and meta-analyses were conducted in accordance with the Preferred Reporting Items for Systematic Reviews and Meta-analyses (PRISMA) Guidelines , 26. We One investigator (YHC) conducted the initial screening of the article titles and abstracts identified in the first screening process. Two investigators (YHC) and (FWT) independently reviewed the full text of the potentially relevant articles for final inclusion, and any disagreement was resolved through discussion.For inclusion, studies had to meet the following criteria: (1) original article; (2) cohort study, case-control study, or nested case-control study; (3) estimates and 95% confidence intervals (CI)  for the association between sleep duration and the incident risk of single common cancer or multiple cancers reported in the literature; (4) adult population; and (5) published in English. Studies were excluded if they (1) had duplicated data; (2) were reviews, reports, clinical trials, or genetic and cell studies; or (3) had insufficient data. If more than one article reported data from a single study, the most recent and complete article was included.Data were extracted by YHC and independently checked by FWT for accuracy and completeness. Any disagreements were resolved by discussion. For each study, we extracted data on study design; first author’s surname; publication year; country of study; study name (cohort studies only); study period; gender of subjects; sample size; types of cancer; number of cases; reference category for sleep; categorization of “short” and “long” sleep duration; fully adjusted relative risks (RRs), odds ratios (ORs), or hazard ratios (HRs) for the associations of both short and long sleep duration with cancer risk; corresponding 95% confidence intervals (95% CIs); and the covariates adjusted in the statistical analysis.Two investigators independently evaluated the quality of the included cohort and case-control studies using the Newcastle–Ottawa Scale (NOS). The selection, comparability, and exposure of each study were broadly assessed and studies were assigned a score from zero to nine. Studies with scores ≥7 were considered of high quality.We performed categorical and dose–response meta-analyses . Random-The categorical meta-analysis was conducted by pooling basic classification results of cancer incidence at different levels of sleep duration. In the original articles, sleep duration was assessed using self-report questionnaire measures of habitual sleep duration. We differentiated three levels of sleep duration: short, medium, and long. Short sleep duration was defined as follows in the different articles (in hours per night): < 5 h , 29, ≤5 Pnonlinearity was identified by testing the null hypothesis that the estimated value of the second spline was equal to zero. If the null hypothesis did not hold, we conducted a linear dose–response meta-analysis to test the cancer risk associated with each additional hour of sleep. Otherwise, a nonlinear dose–response meta-analysis was conducted to identify the cancer risk associated with each hour. For linearity, if a U-, J-, or S-shape curve or a significant nonlinear shape association was observed, we treated the slope as two piecewise linear functions with the cut point of 7 h to show the separate linear trends [Studies with at least three quantitative categories of short or long sleep duration were also included in dose–response analyses. Potential nonlinear dose–response relationships between sleep duration and cancer risk were examined using a restricted cubic splines model with three knots at the 10th, 50th, and 90th percentiles of the distribution. We assigned the median or mean sleep duration in each category to the corresponding OR for each study. If the mean or median duration per category was not reported, the midpoint of the upper and lower boundaries in each category was assigned. For the open-ended risk factor classes , we assigned a value following the algorithms suggested by Il’yasova et al. , choosinr trends . All poox2-value and P-value) and the I2 statistic [P-value of less than 0.1 was considered to indicate the presence of heterogeneity. The I2 statistic was used to test whether the proportion of total variation in the estimates could be explained by heterogeneity rather than chance. I2 values of 25, 50, and 75% were considered evidence of low, moderate, and high heterogeneity, respectively.Heterogeneity among studies was estimated using Cochran’s Q test , gender (women and men), the definition of short or long sleep duration (“≤ 6 or ≤ 5” and “≥ 9 or ≥ 10”), the definition of the reference category (7 h–8 h or 7 h), the definition of short or long sleep duration versus the reference category (“≤6 vs.7” and “≥9 vs. 7”), study quality score (≥7 and < 7) and cancer type to minimize heterogeneity among the included studies. In addition, we estimated the risk of sex hormone-related cancer  , 51. In P-value < 0.05 was considered statistically significant.We visually inspected the symmetry of the funnel plots and performed the Begg regression test and Egger’s linear regression test to assess the possibility of publication bias , 53. ForThe literature searches identified a total of 5288 articles: 1321 from PubMed, 3418 from EMBASE, and 549 from the Cochrane Library , and 1.14 (95% CI: 0.94–1.37) for the 4 case-control studies, with low to moderate heterogeneity  . Exclusion of a single study did not substantially influence the combined estimates for the cohort or case-control studies  in the selected studies  for the 61 cohort studies, with low to moderate heterogeneity  and 1.15 (95% CI: 0.96–1.38) with moderate heterogeneity  for the 4 case-control studies . Exclusion of a single study did not substantially influence the combined estimates for the cohort or case-control studies  in the selected studies  for the 61 cohort studies with low to moderate heterogeneity (P = 0.24) dose–response analysis indicated no relationship between sleep duration and cancer risk  was found, which suggested that increasing sleep duration (in 1-h increments) was not associated with cancer risk . When treated as two piecewise linear functions, among people who slept less than 7 h per night, a 1-h reduction in sleep duration was not associated with an increase in cancer risk , and the results for people with sleep durations of more than 7 h were similar .Sixty-one population-based cohort studies and four case-control studies, including 86,201 cases among 1,550,524 participants, were combined in the dose–response meta-analysis of sleep duration and cancer risk. As shown in Fig. P = 0.005 for nonlinearity, Additional file Using the available data, we examined potential effect modifications by study region, gender, study design (cohort study and case-control study), study quality score, and cancer type. The association was not appreciably modified by study region, gender, study quality, or study design. When stratified by cancer type, we observed a J-shaped curvilinear association, with the lowest colorectal cancer risk at a sleep duration of about 7 h per night  and long sleep duration was associated with increased risk of colorectal cancer (this finding was based on six cohort studies including 8099 incident cases).P = 0.51) between sleep duration and cancer [Most previous reviews and meta-analyses of studies on sleep duration and cancer risk have reported null findings or shown that longer sleep duration might be a risk factor for colorectal cancer, which is consistent with our findings –61. A med cancer  categorid cancer  reportedThe etiology of cancer is multifactorial; genetic, metabolic, environmental, behavioral, and social or cultural factors are major contributors. Although the exact mechanism remains unknown, there are several possible pathways that could explain the association between sleep duration and cancer risk. First, the melatonin hypothesis proposes that shorter sleep duration is associated with decreased levels of melatonin , and melOur findings indicate that long sleep duration is an additional behavioral risk factor for colorectal cancer. The proposed mechanism is that sleep may influence cancer risk via alterations in levels of appetite-regulating hormones, such as leptin and ghrelin , 85, leaWe also performed a subgroup analysis by study population. Short sleep duration was strongly associated with increased cancer incidence in Asian participants but not in American and European participants. Several possible pathways could explain the relationship between short sleep duration and increased cancer incidence in Asians. First, there may be differences in melatonin secretion patterns between Asians and Americans . WetterbOur study addressed the limitations of previous research and had several strengths. First, this meta-analysis was based on up-to-date literature and presented the largest synthesis to date of prospective cohort studies and case-control studies with large sample sizes, which increased the statistical power to detect potential associations. Second, data for the pooled analysis were derived from fully adjusted models in the primary studies, which should reduce the likelihood of confounding. Moreover, the combined use of categorical and dose–response meta-analyses provided more information. Linear and nonlinear relationships were also tested to assess the dose–response relationship, and we performed subgroup and sensitivity analyses on potential confounders. The methodological features of our study enhance the quality of our results and strengthen the validity of the conclusions.Several limitations of this meta-analysis should be considered. First, meta-analyses are greatly influenced by the quality of the individual studies included. Nearly all the included studies relied on self-reports of sleep duration collected via questionnaires or interviews; this type of data may not fully or accurately capture actual sleep duration. Additionally, sleep duration was classified differently across the original studies, and the differences in the reference groups were particularly large. Second, this meta-analysis summarizes results from both prospective cohort and case-control studies. However, there are many differences between these two types of study design, such as different statistical estimates  and different biases. The previous two limitations may have led to heterogeneity in our meta-analysis. Thus, subgroup analyses, sensitivity analyses, and random-effects models were generated to examine sources of heterogeneity. Third, most of the studies assessed sleep duration at a single point in time; this method might not accurately reflect the sustained effects of sleep duration over time when relating them to the long-term development of cancer. Fourth, the available data were limited for several studies, which might have led to a loss of statistical power for some of the subgroup results. In the subgroup analysis of cancer types and study population, the categorical and dose–response analyses produced inconsistent results. It is possible that there was insufficient statistical power to detect a dose–response effect in the different groups; studies with larger subgroup sample sizes are needed to validate these associations. Fifth, most studies reported data for only breast, skin, colorectal, lung, prostate, endometrial, ovarian, and thyroid cancers, except one European study that reported data for all cancer types . TherefoStudy design. Additional targeted biological research is needed to determine the exact mechanisms underlying the association between sleep duration and cancer risk, and population-based long-term epidemiological studies are needed to explore the link between specific (precisely measured) sleep durations and cancer risk.Measurement and assessment of sleep duration. Further epidemiological studies should be conducted using electronic wearable devices to obtain precise, reliable, and scalable objective measures of sleep duration (how long), sleep timing (when), sleep quality (good or bad), and location (longitude and latitude). Such studies should also assess, as directly as possible, the effects of sleep duration and changes in sleep duration on cancer incidence and development during long-term follow-up.Confounding bias. The presence of confounding variables is a limitation of many studies included in this meta-analysis; therefore, future studies should control for the following:The need to sleep with a light on ;Sleep environmental factors ;Sleep quality assessment ;Sleep culture ; andThe use of melatonin supplementationChronobiological variables. Further experimental studies that focus on melatonin, hormonal rhythms, and clock gene expressions are needed to interpret the complex relationships between sleep and cancer.Because of limited evidence and a lack of consensus on the effects of sleep duration on cancer risk, we believe that the following aspects warrant close attention in future investigations of the association between sleep duration and cancer risk , 59, 96:Our findings indicate that neither short nor long sleep duration was significantly associated with cancer risk, although short sleep duration slightly increased cancer risk among Asians and long sleep duration slightly increased the risk of colorectal cancer. Large-scale, well-designed prospective studies are required to further investigate the observed association. Long-term randomized controlled trials are needed to establish causality and to elucidate the mechanisms underlying the association between sleep duration and cancer risk.Additional file 1:PRISMA 2009 Checklist. (DOCX 25 kb)Additional file 2:Characteristics of studies included in the meta-analysis. (DOCX 39 kb)Additional file 3:Study quality of cohort studies included in the analysis of sleep duration and cancer risk. (DOCX 21 kb)Additional file 4:Study quality of case-control studies included in the analysis of sleep duration and cancer risk. (DOCX 20 kb)Additional file 5:Sensitivity analysis of sleep duration and cancer risk, shortest vs. reference analysis. (PDF 194 kb)Additional file 6:Sensitivity analysis of sleep duration and cancer risk, longest vs. reference analysis. (PDF 186 kb)Additional file 7:Association between sleep duration and cancer risk from nonlinear dose–response analysis. (DOCX 20 kb)Additional file 8:Nonlinear dose–response analyses of sleep duration and colorectal cancer risk. (PDF 79 kb)