The etiology of sudden infant death syndrome (SIDS) remains an unsolved problem. The aim of this meta-analysis is to investigate the potential association between monoamine oxidase A (MAOA) promoter variable number tandem repeat (VNTR) polymorphism and SIDS risk.A systematic review and meta-analysis were conducted on studies from accessible electronic databases. Each VNTR variant was examined in each gender independently by comparing with the pooled results of other alleles.A total of six independent case–control studies including 1022 SIDS cases and 1839 controls were enrolled in this meta-analysis. In both of the whole populations and Caucasian populations, male infants with the low-MAOA-expression alleles (2R+3R) were found to exhibit a statistically significant increased risk of SIDS, whereas those with a 4R allele exhibited a reduced risk of SIDS. Besides, an increased risk of SIDS was detected in male Caucasian infants with 2R or 3R alleles. However, none of the allele or genotype variants was associated with SIDS in female victims.In male Caucasian infants, the low expression of MAOA promoter VNTR alleles (2R and 3R) is associated with an increased risk of SIDS, and the existence of the 4R allele could be regarded as a protective factor. Sudden infant death syndrome (SIDS) is defined as the sudden unexpected death that occurs during sleep in infants under one year of age, which cannot be explained by integrated forensic investigations, including systematic autopsy, examination of the circumstances, and review of the clinical history . RegardeSome progress has been made in molecular genetics in research investigating the pathogenic mechanism of SIDS, breaking the impasse and opening new perspectives in this field. Increasing evidence has shown that autonomic nervous system disturbances may play an important role in the occurrence of SIDS, especially those involved in respiratory regulation and arousal reaction. These impaired functions are associated with an abnormal serotonin level in the brainstem, 5-HT1A receptors, and multiple functional gene polymorphisms of the serotonin transporter, leading to a failure in homeostatic defensive responses to life-threatening problems during sleep –11. In aTo further investigate the correlation between neurotransmitter systems and SIDS, a variety of gene polymorphisms have been widely studied , 16, incConsidering that SIDS has been reported to be associated with abnormalities of neurotransmitter systems, MAOA promoter polymorphism, which could regulate the levels of serotonin and noradrenalin, is considered to be of great potential to be correlated with the risk of SIDS. However, no consensus regarding the VNTR polymorphism of the MAOA promoter and the incidence of SIDS has yet been reached in various studies. Filonzi et al.  demonstrA systematic literature search was carried out in several accessible electronic databases, including PubMed, the Cochrane Library, Embase, MEDLINE, Chinese Biological and Medicine Disc, China Science and Technology Journal Database, China National Knowledge Infrastructure, Weipu Database, and Wanfang Database. Because of the language limitation, only literature in English and Chinese was selected. All research published before December 1, 2019, was collected by two independent reviewers (Q.Z. and D.G.) using key words including “sudden infant death syndrome,” “sudden infant death,” “unexpected infant death,” “MAOA,” “MAO-A,” “monoamine oxidase A,” “polymorphism,” and “susceptibility.” There was no restriction on the publication status.Original research regarding the potential association between genotype variants of MAOA gene promoter polymorphism and SIDS susceptibility was taken into consideration. All studies enrolled in the current meta-analysis were selected according to strict inclusion and exclusion criteria.The inclusion criteria were as follows: (1) the studies were originally published as case–control studies; (2) the studies were on the correlation between MAOA gene promoter polymorphism and the risk of SIDS; (3) the infants enrolled in the studies were diagnosed with SIDS according to systematic autopsy and strict guidelines ; and (4) the number of infants with specific gene variants was presented as a real number instead of a statistical analysis result.The exclusion criteria were as follows: (1) reviews, case reports, and case series; (2) non-case–control studies or animal experiments; (3) studies on other diseases or sudden death in adults; (4) duplicate studies; (5) studies without original data or studies reporting the number of genotypes in mixed genders; and (6) studies with unexplained data .Data of the enrolled studies were collected by two independent reviewers (Q.Z. and D.G.) and verified by a third reviewer (Y.Z.) to reach a consensus. Basic information on each study was also collected, including the last name of the first author, year of publication, country and ethnicity of the subjects, methods of genotyping, and the age and number of infants. The quality of each study was assessed using the risk-of-bias tool . Potential bias was also estimated, including random sequence generation, allocation concealment, blinding of participants and personnel, blinding of outcome assessment, incomplete outcome data, selective reporting, and other bias. Bias was assessed and classified into three levels  and visualized in RevMan v.5.8 software (Cochrane Collaboration).I2 was used to assess the heterogeneity between the studies. An I2 value less than 50% (or P > 0.05) suggests significant heterogeneity, suggesting the use of a fixed-effective model to summarize the data. Conversely, an I2 value greater than 50% (or P < 0.05) indicates unacceptable heterogeneity between the studies, highly suggesting the use of a random-effects model. The odds ratio (OR) of the pooled data was calculated using the 95% confidence intervals (CIs), and a Z-test was carried out to determine the significance of the results. A P value of < 0.05 was considered to represent statistical significance.Systematic meta-analysis was carried out using RevMan v.5.8 (Cochrane Collaboration) according to the data extracted from the included studies. To detect the existence of a single allele and the risk of SIDS, each VNTR variant was examined in each gender independently by comparing with the pooled results of other alleles. A comparison between the low-MAOA-expression allele (2R and 3R) and the high-MAOA-expression allele (3.5R and 4R) was also conducted to test the potential correlation between the risk of SIDS and the level of serotonin related to the VNTR variants. It should be noted that the 5R allele was not included in this comparison because of its rare existence in the whole populations and the related unclear serotonin level. In this meta-analysis, Higgins https://www.crd.york.ac.uk/prospero/display_record.php?ID=CRD42020169712).This systematic review and meta-analysis were conducted following the guidelines of preferred reporting items for systematic reviews and meta-analyses (PRISMA). No specific ethical approval from an institutional review board was required for this study as it used data from published literature. This meta-analysis has been prospectively registered on PROSPERO with the identification number of CRD42020169712  and high-MAOA expression alleles (3.5R+4R) . Accordingly, they have successfully examined a total of 261 SIDS cases including 154 male cases and 107 female cases. Considering the possibility of failed PCR examinations, taking the number of all the infants enrolled in their study will obviously lead to some bias. Thus, this inconsistency was considered acceptable and only the number of examined cases and controls were extracted instead of all the enrolled infants. Besides, the numbers of subjects with different alleles and genotypes in control group were calculated by excluding those in the studies of Courts et al. [In total, five variants of MAOA gene promoter polymorphism were estimated in this meta-analysis: 2R, 3R, 3.5R, 4R, and 5R. The number of subjects with different alleles and genotypes is shown in Table s et al.  and Klins et al.  accordinP = .05; Fig. P = 0.03; Fig. Male infants with a 4R allele exhibited a tendency of reduced SIDS risk  and 3R allele . A significant association was also observed in the low-MAOA-expression group, in which the results of 2R and 3R are pooled together  and 5R  did not show a significant correlation to SIDS susceptibility in male Caucasian infants. None of the allele or genotype variants was associated with SIDS in female Caucasian patients.Considering the differences of genotype frequency among different populations, further comparisons were conducted excluding the Caucasian Coriell populations presented in the study of Paterson et al.  as most P = 0.04). The stability of other comparisons conducted on male infants was not affected by removing any of the studies.The effect of each individual study on the total results was estimated by removing the results of the studies one by one to observe the change that occurs in the overall results. After excluding the study of Gross et al. , the stuIt has been reported in previous research so-called “triple-risk-hypothesis” that the development of SIDS is affected by the interaction between environmental challenges and the diverse genetic properties of the individuals as well as the developmental factors . There wTo the authors’ knowledge, this meta-analysis is the first meta-analysis aiming to assess the potential correlation between MAOA promoter VNTR polymorphism and the risk of SIDS, and it is also the first meta-analysis discussing the association between gene polymorphisms and susceptibility to SIDS. By pooling the results of published studies with different sample sizes and applying subgroup analysis, reliable evidence was provided for the results. It was observed that the 2R and 3R alleles increased the risk of SIDS in Caucasian male infants, and that the presence of 4R decreases the risk of SIDS in male infants, whether Caucasian or mixed populations, which can be regarded as a protective factor. It was also confirmed that there is a correlation between the low-MAOA-expression allelic variants  and the development of SIDS in male infants, indicating potential correlations between low serotonin levels and the development of SIDS, which also coincided with previously published studies. However, it is worth mentioning that the study of Gross et al. , which wPreviously published studies have proposed a subset of SIDS with a high serum level of serotonin, which may be related to 5-HT network abnormalities, suggesting that an increased serum serotonin level is a potential biomarker for the diagnosis of SIDS in forensic science . A mechaBesides, male infants were reported to exhibit a 30–50% higher risk of SIDS comparing to female infants , 34, witGiven the triple-risk-hypothesis, SIDS risk has been proved to be correlated with quite different genetic predispositions because different evidence could be integrated into various hypotheses for explanation of the SIDS etiology. Thus, gene polymorphisms like MAOA locus, which may not constitute a lethal factor, are better to be regarded as a role disposes to SIDS and added to the genetic background influence in a triple-risk model. Limited by the available information in the enrolled studies, heterogeneity raised by environmental and developmental factors such as age, temperature, sleeping posture, and feeding patterns could not be excluded, although it did not show up in this current meta-analysis. To get a better grasp of the possible etiology, studies with more detailed information will be encouraged in the future.This meta-analysis included some limitations that should be addressed here. First of all, given that only individuals of the same race were included in the studies, associations could be examined only in Caucasian infants, and thus studies on populations with other ethnicities are highly recommended in the future. However, this limitation might have been useful for the current study because it helped exclude the influence of hybrid genes introduced by mixed populations, which also exhibited low heterogeneity between the studies in the meta-analysis. Second, publication bias tends to exist in research, although none was observed in this study. Except for unpublished studies with negative results, it is possible that more studies published in other languages exist. Third, the number of studies included in the current meta-analysis is relatively small compared with other studies, and there are more published studies investigating the correlation between the polymorphism of MAOA promoter and sudden infant death risk. However, this study is designed with strict inclusion and exclusion criteria in the process of study selection. Firstly, considering that the MAOA gene is an X-chromosome located gene, the studies presented in mixed-gender would obviously introduce some potential bias as the frequency of the female subjects could be counted twice. Secondly, there were odd numbers presented in some studies regarding the total alleles of the female populations, which we regarded them as “unexplained” and were further excluded due to unreliable data. Thirdly, some studies reported the updated data based on their previously published work, which included repeated populations. Thus, only the latest study or the study with the largest populations was included in this current work. In order to ensure the accuracy and scientificity of the meta-analysis, studies with the problems mentioned above have to be excluded. To our knowledge, there were no more published studies could be included. We believe it is necessary to complete the study with strict inclusion criteria, and we hope this meta-analysis could provide a reference for further researches to avoid the problems mentioned above. Besides, as gene polymorphism may combine with various factors and contribute to the development of SIDS, it is very important to perform further subgroup analyses. However, this was not possible here because of the restrictions due to the limited information of the subjects in the included studies. Thus, it is highly recommended to perform more systematically designed studies with detailed and scientifically presented data in the future.According to this meta-analysis, low-expression MAOA promoter VNTR alleles (2R and 3R) are associated with an increased risk of SIDS compared to high-expression alleles (3.5R and 4R) in male Caucasian infants, and the existence of the 4R allele could be regarded as a protective factor against the development of SIDS in male Caucasian infants. This study also suggested that abnormalities in the serotonergic system and nervous system disturbances may contribute to the development of SIDS. In conclusion, it is recommended to perform more systematically designed studies with detailed data in the future.


Objective: Currently available evidence regarding the association between collagen type XII α1 (COL12A1) polymorphism and risk of anterior cruciate ligament rupture (ACLR) remains elusive. The aim of our present study was to assess the association between COL12A1 rs970547 polymorphism and ACLR risk.Methods: Five online databases, namely, PubMed, EMBASE, ISI Web of Science, CENTRAL, and CNKI, were searched from their inception data up to December 2020 to identify relative observational studies. The methodological quality of each individual study was evaluated using the Newcastle-Ottawa Scale (NOS). The “model-free approach” was employed to estimate the magnitude of effect of COL12A1 rs970547 polymorphism on ACLR, and the association was expressed using odds ratio (OR) and its associated 95% confidence interval (95% CI). Subgroup analysis was performed by ethnicity and sex of included subjects.Results: Eight studies involving 1,477 subjects with ACLR and 100,439 healthy controls were finally included in our study. The methodological quality of included studies was deemed moderate to high based on NOS scores. The “model-free” approach suggested no genotype differences between ACLR and healthy control for the rs970547 polymorphism, but we still used the allele model to present the combined data. Under the random-effect model, there was no significant difference in the frequency of effecting allele between ACLR and control . Stratified analysis by sex and ethnicity also showed no difference in allele frequency.Conclusion: The findings of this current meta-analysis suggested that rs970547 was not associated with ACLR risk in male, female, and the overall population among Asians or Caucasians. The anterior cruciate ligament (ACL) is the primary stabilizer in the human knee joint, which acts by antagonizing anterior translation of the tibia on the femur and counteracting rotational and valgus stresses. With the popularity and extensive development of national sports, the incidence of the anterior cruciate ligament rupture (ACLR) is in an increasing trend. According to epidemiological reports, ACLR accounts for as much as 50% of all knee injuries and some authors made an estimation of 250,000 ACL injuries per year worldwide . In immunoelectron microscopy, collagen XII is associated with the surface of collagen fibrils where it is able to form interfibrillar connections and mediate fibril interaction with other extracellular and cell surface molecules within tendons and other tissues  in exonic regions of COL12A1 have been identified and two SNPs (rs240736 and rs970547) are considered as non-synonymous SNPs, indicating that amino acid sequence could possibly change due to these two mutations. Up to now, several studies focusing on rs970547 SNP and the susceptibility of ACLR have been reported and only few studies have been featured to assess the association between rs240736 polymorphism and risk of ACLR. However, the results were mixed and inconclusive due to clinical heterogeneity, different ethnic populations or sexes, and small sample sizes. To overcome the limitations of these studies for elucidating the genetic roles for rs970547 SNP in the risk of ACLR, we performed a comprehensive meta-analysis of identified studies to evaluate the contradictory results from these relevant studies and to clarify the associations between rs970547 SNP in COL12A1 gene and ACLR susceptibility.This meta-analysis was conducted on the basis of the PRISMA guideline . The reference lists of relative systematic reviews and meta-analyses were also checked by two independent reviewers for additional relative studies.A systematic literature search was performed using five online databases, namely, PubMed, EMBASE, ISI Web of Science, CENTRAL, and CNKI, to identify observational studies that assessed the relationship between COL12A1 polymorphism and risk of ACLR. No language restriction was imposed on the literature search; studies published before December 2020 were searched. The search strings for English databases were as follows: “COL12A1 or collagen type XII α1 or rs970547” and “polymorphism or variant or single nucleotide polymorphism or SNP or SNPs” and “anterior cruciate ligament rupture or ACLR or ACL or anterior cruciate ligament tear or anterior cruciate ligament injury.” For Chinese database, we used “Qian Jiao Cha Ren Dai” and “COL12A1” and “Duo Tai Xing” to retrieve relative studies. Effort was also made to avoid missing Genome-wide Association Study (GWAS) that did not use COL12A1 as part of keywords or abstract; we also searched the GWAS Catalog  principle: (i) cases were patients that surgically confirmed with ACLR; (ii) the variant of interest in our current study was rs970547 in COL12A1 gene; (iii) the control subjects were healthy and unrelated individuals without any self-reported history of ACL injury; (iv) the only outcome of our study was the association between COL12A1 polymorphism and risk of ACLR, as indicated by crude odds ratio (OR) and its associated 95% confidence interval (95% CI). The ORs could be either reported in each included study or calculated based on the genotypes distribution in ACLR and healthy control groups; (v) study type was observational study on humans .On the other hand, studies would be excluded if they met the following exclusion criteria: (i) animal studies and cellular experiments; (ii) reviews, editorials, case series, or case report. In case that several studies reported overlapping data, only the most comprehensive one would be considered for inclusion.Data were extracted using a predetermined data collection sheet that included the surname of the first author, year of publication, country of origin, ethnicity of enrolled subjects, number of subjects in ACLR and control group, genotyping method used in each study, genotype counts in ACLR groups and control groups, and Hardy–Weinberg Equilibrium (HWE) results for control subjects. Discrepancies occurred during the data collection process were resolved through discussion between two investigators (ZL and WW). Otherwise, the third investigator (JH) would be consulted for his opinion.Two reviewers (ZL and WW) independently assessed the methodological quality of each individual study using the Newcastle-Ottawa Scale (NOS)  Recessive model: if OR1 = OR3 ≠ 1 and OR2 = 1;(2) Dominant model: if OR1 = OR2 ≠ 1 and OR3 = 1;(3) Complete over-dominant model: if OR1 = 1, OR2 = 1/OR3 ≠ 1;(4) Codominant model: if OR1 > OR2 > 1 and OR1 > OR3 > 1 (or OR1 < OR2 <1 and OR1 < OR3 <1).Q-statistical test and I2 test, where p < 0.1 and I2 > 50% indicated statistically significant heterogeneity .In order to evaluate whether our included studies had an adequate sample size to detect 30% increased or decreased risk of ACLR, we averaged the minor allele frequency (MAF) among control subjects. It turned out that the MAF among Asians and Caucasians were 38.13 and 22.54%, respectively. Therefore, at least 294 (147 cases and 147 controls) and 658 (329 cases and 329 controls) subjects were required for studies among Asians and Caucasians, respectively, to detect the association between the rs970547 polymorphism and susceptibility to ACLR, with a level of significance of 0.05 and power of 0.8  and OR3  was nonsignificant while only OR2  was marginally significant. Based on the relationships between these three pairwise comparisons, rs970547 was not significantly associated with risk of ACLR. Therefore, we decided to illustrate the meta-analysis results using the allele model. As shown in p = 0.28) when using the allele model, and statistically significant heterogeneity was observed . Random-effect model was used due to anticipated presence of heterogeneity.Prior to combining data from each individual study, we applied a “model-free approach” to estimate the genetic effect of rs970547 on risk of ACLR. Seven studies . Similarly, no association was found among females . Subgroup analysis by ethnicity was also performed to test the genetic effect of rs970547 in different populations. No association of rs970547 and ACLR risk could be found among either Asians  or Caucasians , as shown in Subgroup analysis was performed according to ethnicity and sex. Five studies  to low . The funnel plot  and Egger's test  also suggested no statistically significant publication bias.The robustness of our conclusion was further confirmed by the leave-one-out sensitivity analysis (detailed data not shown). It should be noted that, after the removal of Kang's study, the intra-study heterogeneity diminished from moderate  and found no data to support the association between rs970547 and altered gene expression in different human tissues. The results of our current study also suggested no association between rs970547 and risk of ACLR. In our included studies, the male and female in control and ACLR groups were not similarly matched in physical activity. Briefly, the female in the ACLR group had significantly longer period of non-contact jumping sports while the male had longer history of contact sports . Secondly, although we performed stratification analyses by ethnicity or sex, we were unable to rule out the intrinsic and extrinsic factors mixed in study. For instance, the current body of evidence did not allow us to perform a subgroup analysis of injury mechanisms (contact or non-contact mechanism). Thirdly, majority of our included studies were conducted in a Caucasian population. Only three studies were designed to investigate the genetic effect of COL12A1 polymorphisms in an Asian population, and the sample size was limited. More high-quality studies within different ethnic backgrounds are needed to verify the genetic effect of COL12A1 polymorphisms in risk of ACLR. If the effect of rs970547 in pathogenesis of ACLR could be confirmed by future studies, it would be of great interest to dive deeper into the possible link of COL12A1 polymorphisms and other types of other musculoskeletal soft-tissue injuries, such as Achilles tendon injuries and rotator cuff injuries, to facilitate our understanding of the biomechanical properties of tendons at a molecular level as well as ligament and tendon injuries that occur during physical activities.Taken together, the findings of this current meta-analysis suggested that rs970547 was not associated with ACLR risk in male, female, and the overall population among Asians or Caucasians. However, larger-scale studies within different ethnic backgrounds are still warranted to confirm the findings of our study.The original contributions presented in the study are included in the article/supplementary material, further inquiries can be directed to the corresponding authors.J-mH and Z-tL conceived and devised the study. J-mH, Z-tL, and D-mZ performed the experiments and wrote the paper. WW and D-mZ analyzed the data. WW and Z-tL revised the manuscript. All authors have contributed to the final version and approved the publication of the final manuscript.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.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.


Despite the recognition of the importance of indoor microbial exposures on children’s health, the role of different microbial agents in development and aggravation of respiratory symptoms and diseases is only poorly understood. This study aimed to assess whether exposure to microbial aerosols within the indoor environment are associated with respiratory symptoms among children under-5 years of age.I2 statistics using RevMan 5.3.A systematic literature search was conducted on PubMed, Web of Science, GreenFILE, ScienceDirect, EMBASE and Cochrane library through February 2020. Studies that investigated the exposure–response relationship between components of the indoor microbial communities and respiratory symptoms among under-five children were eligible for inclusion. A random-effect meta-analysis was applied to estimate pooled relative risk (RR) and 95% confidence interval (CI) for study specific high versus low microbial exposures. The potential effect of individual studies on the overall estimate was evaluated using leave-one-out analysis, while heterogeneity was evaluated by P = 0.001]. The association was strongest with exposure to a combination of Aspergillus, Penicillium, Cladosporium and Alternaria species . Stratified analysis revealed an increased risk of wheeze  from any microbial exposure.Fifteen studies were eligible for inclusion in a meta-analysis. The pooled risk estimate suggested that increased microbial exposure was associated with an increased risk of respiratory symptoms [pooled relative risk (RR): 1.24 , Microbial exposures are, in general, associated with risk of respiratory symptoms. Future studies are needed to study the indoor microbiome more comprehensively, and to investigate the mechanism of these associations.The online version contains supplementary material available at 10.1186/s12940-021-00759-2. The microbial community within indoor environments such as dwellings where humans, especially under-five children, spend more than 90% of their time, consists of a wide range of microorganisms including bacteria, fungi, and viruses –4. The mMost epidemiological studies have been heterogeneous in assessing different respiratory health effects associated with exposure to specific microbial components, especially endotoxins, and culturable molds –17 whileRespiratory health effects have been the subject of recent research among preschool children , 26 and AND “damp” OR “fungi” OR “mold” OR “mould” OR “fungal” OR “fungus” OR “bacteria” OR “virus” OR “microbial” OR “microbiome” OR “microbial diversity” OR “microbial load” OR “microbial burden” OR “microbiota” OR “biodiversity” AND “respiratory symptoms” OR “allergy” OR “hay fever” OR “cough” OR “fever” OR “difficulty breathing” OR “wheeze” OR “allergic rhinitis” OR “sinusitis” OR “asthma”. Title and abstract of each article was evaluated independently and differences on which publication(s) to include were clarified by recourse to a third reviewer. A snowball search was also carried out by screening reference lists of publications and reviews.The protocol for this meta-analysis was reported using MOOSE guideline  and regiStudies eligible for inclusion in the meta-analysis were epidemiological reports in humans with exposure–response relationship between indoor microbial aerosols and respiratory symptoms. Inclusion criteria were: (I) qualitative or quantitative assessment of indoor microbial aerosols. Studies that reported exposure to specific microbial communities  or presence/absence of visible molds were included; (II) studies conducted among children aged ≤ 5 years; (III) respiratory symptoms  and/or asthma adequately defined and described; (IV) full text articles in English originally published in peer reviewed journals. Literature reviews, abstracts, letters to the editor, case reports, and non-human studies were excluded.Assessment of articles was performed in EndNote databases. All duplicates were removed and studies were selected based on title or abstract for full text-screening. For studies that were excluded, the reasons for exclusion were listed.From the studies that met the inclusion criteria, the following information was extracted: (1) name and initials of the first author, (2) year of publication, (3) country, (4) type of sample, (5) sampling equipment, (6) microbial agent(s), (7) analytical method, (8) level and measure of exposure, (9) respiratory outcome definition, (10) number of cases, (11) sample size and (12) effect estimate. All incongruities from the data extracted were resolved by a third author.Two team members (AGF and NJ) assessed the methodological quality and risk of bias of the included studies using the Cochrane Collaboration guidelines  and the All statistical analysis was carried out using Review Manager 5.3. We applied the inverse of variance method for weighting and computed the summary effect estimates by first log-transforming all relative risk (RR) and 95% confidence interval (CI) for high vs low category (as reference) of microbial exposure. The standard error (SE) was estimated based on the formula:I2 test statistics. Where I2 statistics ≥ 50%, a random effect model was employed, otherwise, a fixed effect model , P < 0.0001; I2 = 0% , P < 0.00001; I2 = 0%] (model 1). Three studies  (model 2). The study-specific estimates based on two studies  (model 3) ; TFC and wheeze (B); Aspergillus spp. and wheeze (C); Penicillium spp. and wheeze (D); Cladosporium spp and wheeze (E); Endotoxin and wheeze (F). Figure S2. Funnel plots with pseudo-95% confidence limit for any IM exposure and allergic rhinitis among U-5C (A); visible mold and allergic rhinitis (B); Endotoxin and allergic rhinitis (C). Figure S3. Funnel plots with pseudo-95% confidence limit for any IM exposure/endotoxin and asthma among U-5C.


The consumption of legumes is promoted as part of a healthy diet in many countries but associations of total and types of legume consumption with type 2 diabetes (T2D) are not well established. Analyses across diverse populations are lacking despite the availability of unpublished legume consumption data in prospective cohort studies.To examine the prospective associations of total and types of legume intake with the risk of incident T2D.Meta-analyses of associations between total legume, pulse, and soy consumption and T2D were conducted using a federated approach without physical data-pooling. Prospective cohorts were included if legume exposure and T2D outcome data were available and the cohort investigators agreed to participate. We estimated incidence rate ratios (IRRs) and CIs of associations using individual participant data including ≤42,473 incident cases among 807,785 adults without diabetes in 27 cohorts across the Americas, Eastern Mediterranean, Europe, and Western Pacific. Random-effects meta-analysis was used to combine effect estimates and estimate heterogeneity.I2 = 74%). Analysis by region showed no evidence of associations in the Americas, Eastern Mediterranean, and Western Pacific. The positive association in Europe  was mainly driven by studies from Germany, UK, and Sweden. No evidence of associations was observed for the consumption of pulses or soy.Median total legume intake ranged from 0–140 g/d across cohorts. We observed a weak positive association between total legume consumption and T2D  per 20 g/d higher intake, with moderately high heterogeneity (These findings suggest no evidence of an association of legume intakes with T2D in several world regions. The positive association observed in some European studies warrants further investigation relating to overall dietary contexts in which legumes are consumed, including accompanying foods which may be positively associated with T2D. Type 2 diabetes (T2D) is a major worldwide public health issue which decreases both quality of life and life expectancy , with prStudying the association between legume consumption and T2D is complicated by inconsistency in the definition of legumes and legume subtypes. Legumes are defined as flowering plants in the Leguminosae botanical family, with the fruit enclosed in a pod; they include pulses such as beans, lentils, peas, and chickpeas , plus sohttp://www.interconnect-diabetes.eu/data-discovery/) to identify cohorts potentially suitable for inclusion in the legumes collaborative group. The InterConnect registry was compiled using systematic searches of the literature alongside surveys of other online study registries, surveys of websites relating to consortia of studies, and searches of the gray literature to identify unpublished data. We attempted to contact a total of 103 cohorts , among which we were unable to establish contact with 46, a further 20 did not have sufficient data on exposure, outcome, or covariates, 4 stated they had no capacity to contribute, 4 declared no interest, and 2 cohorts had only recently started recruitment .InterConnect aims to optimize the use of existing individual participant data by enabling cross-cohort analyses within consortia without the pooling of data at a central location. InterConnect uses a federated meta-analysis approach  and aimsSupplemental Table 2). Data for 8 cohorts were obtained by approval of data sharing requests, whereas the remaining 19 cohorts uploaded data to a server to allow federated analysis. We classified regions according to the WHO . The majority of cohorts provided exposure data in the format of grams per day. For those that did not, variable-specific standard portion sizes sourced from the USDA .Of the 27 cohorts, 17 (63% of cohorts) used semiquantitative FFQs, 4 (15%) cohorts used a quantitative dietary questionnaire, 3 (11%) cohorts used an interviewer-administered dietary history, 2 (7%) cohorts used a 24-h recall, and 1 (4%) cohort used either an FFQ or a quantitative dietary questionnaire depending on location  for peas, then the pulse variable would be set to missing for that participant as pulse includes both lentils and peas. However, if >10% of participants in a cohort had missing values for a constituent food variable, we consulted the host institution to ascertain whether data were truly missing or should be set to zero values. If all constituent food variables were missing, then the summed variable was set to missing.1) ascertained by linkage to a registry or medical record; 2) confirmed antidiabetic medication usage; 3) self-report of physician diagnosis or antidiabetic medication, verified by any of the following: a) ≥1 additional source from 1 or 2 above, b) biochemical measurement (glucose or glycated hemoglobin), c) a validation study with high concordance. For the secondary outcome, which was more inclusive, a case of incident T2D was confirmed by any of the following criteria: 1) ascertained by linkage to a registry or medical record; 2) confirmed antidiabetic medication usage; 3) self-report of physician diagnosis or antidiabetic medication; or 4) biochemical measurement (glucose or glycated hemoglobin).To minimize heterogeneity resulting from variation in T2D diagnosis across cohorts, we created 2 harmonized outcomes: the primary outcome was defined as “clinically incident T2D” and the secondary outcome was defined as “incident T2D.” For the primary outcome, a confirmed clinical case of incident T2D was considered as fulfilling any 1 or more of the following criteria: Federated analyses were conducted using R (R Core Team) within the DataSHIELD federated meta-analysis programming library. DataSHIELD permits analyses to be undertaken without the necessity for individual participant data to be transferred and stored at a central location . InsteadSupplemental Tables 5 and 6).For the main analyses, we excluded participants with a diagnosis of diabetes at baseline, those reporting extreme energy intakes  , and thoI2 statistic) using STATA/SE 14.2 (StataCorp).Incidence rate ratios (IRRs) and 95% CI for T2D according to total and types of legume intake were estimated in each individual cohort using piecewise Poisson regression, which is available in the DataShield R programming library as a close approximation of the Cox model . For anan method . Random-Supplemental Table 7), we fitted 4 models: Model 1 adjusted for sociodemographic characteristics and lifestyle behavioral covariates ; Model 2 additionally included BMI; Model 3 was as Model 2 plus prevalent baseline comorbidities ; Model 4 was as Model 3 plus the consumption of red and processed meat, fruit, vegetables, sugary beverages, dairy products, and fish. Due to limited data on family history of diabetes and waist circumference, we conducted additional analyses fitting those covariates in the subgroup of participants for whom the information was available.To assess whether results varied by adjusting for different sets of covariates  by adding a product term between legume intake and each of these variables separately to the regression analyses, with subsequent stratification if the P-interaction <0.05. Pooled effect sizes and I2 statistics (if applicable) were also presented for each geographic region represented . To assess sources of heterogeneity between cohorts, meta-regression (if appropriate) was performed by regressing effect estimates on median intakes of legumes, geographic region, and dietary assessment method.Tests for multiplicative interaction were performed in Model 4 for each of sex, age, and categories of BMI . This was conducted only in EPIC-InterAct due to the availability of additional data and the presence of outlying effect sizes. We also examined the potential for reverse causality by estimating associations with exclusion of T2D cases occurring in the first 2 y of follow-up. Due to large numbers of participants not consuming any legumes , we also estimated IRR in only those reporting total legume consumption >0 g/d. We investigated whether the high degree of heterogeneity in Europe was explained by effect sizes in Whitehall II, EPIC-InterAct Sweden, and EPIC-InterAct Germany by repeating the primary analysis with the omission of these studies.The collaborative group of 27 cohorts included a total of 807,785 individuals (I2 = 74%). Analysis by region showed no evidence of associations in the Americas , Eastern Mediterranean , and Western Pacific . A positive association was observed in Europe where heterogeneity was high . The overall association for total legume corresponded to IRR = 1.17 (95% CI: 1.04 to 1.32) if scaled to 140 g/d higher consumption . No evidence for overall or region-specific associations was observed for pulses .Overall, there was a weak positive association between total legume consumption and T2D  per 20 g/d higher intake  (P-interaction >0.05). In the subset of cohorts for which we were able to additionally adjust for family history of diabetes and waist circumference , there were only minor changes in the observed associations between total legume intake and incident T2D , median consumption , and method of dietary assessment  did not appear to be strong predictors for the heterogeneity in effect size between cohorts.There was no evidence of interactions of total or types of legume consumption with sex, age, or BMI . A positive association was also observed when excluding participants reporting zero consumption of legumes thus restricting the analysis to consumers only . To investigate potential residual confounding, we entered additional covariates to the model in a subset of cohorts with available data. Further adjustment did not meaningfully alter individual or pooled effect sizes, including the observed positive associations between legume intake and T2D in the EPIC-InterAct studies for Germany and Sweden . Omitting Whitehall II, EPIC-InterAct Sweden, and EPIC-InterAct Germany from the primary analysis resulted in a reduction of the I2 value from 82% to 0% in Europe. The I2 value for the overall pooled result including all regions reduced from 74% to 17%.To explore the possibility of reverse causality, we repeated the analysis excluding T2D events in the first 2 y of follow-up and observed a positive association between total legume consumption and T2D, similar to the primary analysis , but there was a modest positive association in Europe. We found no evidence of associations of any of the types of legumes (pulses or soy) with T2D in any of the 4 world regions.I2 = 85%) (I2 = 74%). However, because we were able to include data from 27 studies across different world regions, including previously unpublished studies, we identified that a positive association was evident in Europe, whereas no association was observed in the other regions represented. Similarly, the lack of an association that we observed for total soy intake and T2D in the current analysis was consistent with the prior meta-analysis (I2 = 57%) than the prior meta-analysis (I2 = 91%).The current study builds upon a recent systematic review and meta-analysis of 7 published studies which reported no evidence of an association between total legume intake and T2D, with evidence of heterogeneity (2 = 85%) . The preanalysis  althoughI2 = 82%) and the association appeared to be driven by a small number of cohorts, specifically Whitehall II (UK), EPIC-InterAct Germany, and EPIC-InterAct Sweden. Previous analysis in the EPIC-Potsdam Study reported that legume intake was associated with a dietary pattern linked to greater risk of T2D, and the authors discussed that this may be because of legume consumption in mixed dishes (e.g. in stews) that include pork or beef  emphasizing the need to better understand population-specific associations with T2D.We observed no evidence of associations of total legume consumption with T2D risk in the American continent, consistent with previous findings , 18. In We have contributed new evidence in the Western Pacific region, additional to reports from 2 previous studies. Villegas et al.  reportedI2 = 75%), mean sample size was small (n = 23), and mean consumption (152 g/d) was higher than any cohort described here  for an analysis of legume consumption and T2D. Previous studies included between 266 and 4529 incident T2D cases, whereas a meta-analysis reporting no overall association included 7 published studies with a total of 11,232 cases (A key strength of this work was the analysis of individual participant data from 27 cohorts, contributing the largest number of confirmed incident T2D cases to date (32 cases . The fedThis study has a number of limitations. There was variation in both the portion sizes used by each cohort and those which we assigned at analysis. The accuracy of these portion sizes may vary according to the type of legumes in each cohort; for example, the legume content of bean soup may be exaggerated by the portion size assigned. Dietary assessment is complex and prone to measurement error, and legume consumption is particularly difficult to capture consistently across different populations because of the wide variety of legume types and subtypes, and how they are prepared, cooked, and consumed, which we were unable to examine. Specific validity of legume intake from dietary assessment instruments is generally limited in the published literature and was not available for the current research. Although we made considerable efforts to harmonize the exposure variables, heterogeneity resulting from dietary assessment cannot be ruled out, and it may be necessary to examine legume intake with greater specificity. We used diet data measured only at baseline but recognize that intraindividual variation over time might be present which may bias our findings in an unknown direction. However, most published studies have used solely baseline dietary data , and somIn summary, legume consumption appeared to have a null association with T2D in most of the world regions we investigated, including those where they are consumed as staple foods. The modest positive associations observed in some European cohorts require further investigation to understand the underlying reasons including cooking methods as well as accompanying foods and overall dietary patterns. Until findings from such further research are available, individuals and health professionals should continue to follow existing regional or other dietary guidelines.nxaa447_Supplemental_FileClick here for additional data file.


Most previous studies compared the risk for non-melanoma skin cancer (NMSC) in biologic-treated common inflammatory diseases with the general population. Whether the increased NMSC risk is caused by the disease itself, the biologics, or both remains unknown.We systematically searched PubMed, Embase, Medline, Web of Science, and Cochrane Library from inception to May 2021. Studies were included if they assessed the risk of NMSC for rheumatoid arthritis (RA), inflammatory bowel disease (IBD), or psoriasis patients treated with biologics compared with patients not receiving biologics. Pooled relative risks (RRs) and 95% confidence intervals (CIs) were calculated using the fixed- or random-effects model.The current meta-analysis included 12 studies. Compared with patients with the inflammatory disease without biologics, patients receiving biological therapy were associated with an increased risk for NMSC , especially in patients with RA  and psoriasis , but not in patients with IBD . The risks for squamous cell skin cancer and basal cell skin cancer were both increased for patients receiving biologics. However, the risk of NMSC did not increase in patients treated with biologics less than 2 years.Current evidence suggests that increased risk of NMSC was identified in RA and psoriasis treated with biologics compared with patients not receiving biologics, but not in patients with IBD. The inner cause for the increased risk of NMSC in IBD patients should be further discussed.The online version contains supplementary material available at 10.1186/s12935-021-02325-9. Rheumatoid arthritis (RA), inflammatory bowel disease (IBD), and psoriasis are three common immune-mediated inflammatory diseases involved with epithelial or connective tissue with overlapping genetic susceptibility and relatively high incidence –4. PreviEsse et al. identified no significant association between the risk of melanoma and biological treatment for patients with common inflammatory diseases compared with those receiving non-biological therapy, which showed that biological therapy is not critical in developing melanoma for patients with inflammatory diseases . Van LümPrevious studies in biologic-treated IBD and psoriasis have found an elevated incidence of NMSC in pan-cancer research –20, whilTo address the above issues, we performed a systematic review evaluating the risk of NMSC in patients with common inflammatory diseases treated with biologics. Considering the inherent relevance of the three diseases, connective studies are necessary to clarify whether the source of the increased risk of NMSC is related to biological agents, the diseases, or both. We aimed to present a therapeutically meaningful review of the available information to help clinicians make better therapy decisions.The following terms were used to search PubMed, Embase, Medline, Web of Science, and Cochrane Library from their creation to May 2021 for eligible studies, with no language constraints: “arthritis rheumatoid” or “rheumatoid arthritis” or “rheumatoid chronic arthritis” or “inflammatory bowel disease*” or “ulcerative colitis” or “crohn” or “psoriasis” or “inflammatory disease*” or “immune-mediated disease*,” “skin cancer*” or “skin neoplasm*” or “NMSC,” and “TNFI” or “tumor necrosis factor inhibitor*” or “tumor necrosis factor-α antagonist*” or “TNF-α inhibitor*” or “anti-TNF” or “biologic*” or “infliximab” or “adalimumab” or “etanercept” or “golimumab” or “certolizumab” or “ustekinumab” or “rituximab” or “abatacept” or “tocilizumab” or “natalizumab” or “vedolizumab.” The details of the search strategy are shown in Additional file Randomized clinical trials, cohort studies, and nested case–control studies to investigate the risk of NMSC in patients with RA, IBD or psoriasis were included for further analysis. Studies were deemed potentially eligible if they matched the preset criteria listed below: (1) Studies on people with RA, IBD, or psoriasis; (2) treatment based on biologics; and (3) the risk estimates and 95% CI of NMSC connected with biologics compared with those not receiving biologics. Studies were excluded if they met the following criteria: (1) use of the general population as the comparator; (2) non-clinical studies, such as animal studies; and (3) no relative risk could be extracted. Two researchers independently assessed study eligibility by screening study titles and abstracts and then reading the studies in total. Discrepancies about eligibility were settled by consensus with the third investigator.Data extraction was performed independently by two reviewers. The following information was extracted from each paper: lead author; publication year; where the study was conducted; data source; study design; types of biological therapy; comparator therapy; treatment duration; disease severity indicators; sample size; effect size data and associated 95% CI; and adjustment variables. The Newcastle–Ottawa Quality Assessment Scale (NOS) was used to evaluate study selection, matching, and outcome of the included studies –32.2 statistic was used to quantify it. An I2 score of 50% or higher was considered to show significant heterogeneity. In anticipation of clinical heterogeneity, the random-effects model was performed. Begg’s and Egger’s tests were used to investigate publication bias. Subgroup analyses were conducted based on types of NMSC, types of biological therapy, study quality, treatment years, and sample size. All statistical analyses were conducted using Stata statistical software, version 15.1 .The relative risk reported in included articles were chosen for inclusion in the meta-analysis. We calculated summary RRs and associated 95% CIs for all outcomes using random- or fixed-effects models . In studThe process of article selection is shown in Fig. The 12 included studies were published from 2007 to 2019, and all were observational studies conducted in the USA (n = 8), Sweden (n = 2), the UK (n = 1), and Demark  and psoriasis (I2 = 0.0%) subgroups. No publication bias was indicated in the included studies , hydroxychloroquine, sulfasalazine, or leflunomide for RA patients and MTX, cyclosporine, ciclosporin, hydroxyurea, mycophenolate mofetil, sulfasalazine, or thioguanine for psoriasis patients , 44. TheTo the best of our knowledge, this work is the first meta-analysis to explicitly investigate the risk of NMSC in patients with IBD and psoriasis who received biological therapy compared with their biologic-naïve patients. A meta-analysis of any cancer reported an increased risk of NMSC in patients with IBD without assessing the effect of any treatments . At presAt present, the only systemic review of the pan-cancer study involving the risk of NMSC in patients with psoriasis also treated the general population as comparator group without estimating relevant effect, which demonstrated an increased risk of NMSC in patients with biologic-treated psoriasis compared with the general population . In the Previous meta-analyses have indicated a relationship between the risk of NMSC and biological treatment in RA patients, which yielded similar results to our study. Xie et al. summarized four studies and demonstrated that patients with biologic-treated RA had a higher risk of NMSC  compared with non-biologic treated individuals . The datBCC (about 70%) and SCC (about 25%) are the most prevalent kinds of NMSC . Only twIn subgroup analysis, biological therapy for more than 2 years was significantly associated with the increased risk of NMSC in three inflammatory diseases, while no significant association was found for less than 2 years’ treatment and the increased risk of NMSC, indicating that the risk of NMSC was related to the duration of biological therapy. However, not all the included studies contained the information of mean treatment duration. For IBD patients, the data of mean treatment duration were missed, and 3 months was used as the minimal treatment duration , which sTo further clarify the impact of different biological treatments on the occurrence of NMSC, we conducted a subgroup analysis by different biological treatments. The results showed that TNFIs and abatacept were both associated with the increased risk of NMSC in inflammatory diseases. However, no significant difference was identified between the use of rituximab and the risk of NMSC in RA patients.TNFIs were the most often used biologics in all the three common inflammatory diseases groups, as shown in Table As a cytotoxic T-lymphocyte-associated protein 4-fusion protein, abatacept specifically inhibits T cell activation, which has been licensed to treat RA. Abatacept is likely related to an increased risk of cancer as it reduces anti-tumor response and immune surveillance , 59. In The severity of the disease could also be an important factor in analyzing the risk of NMSC . PreviouThe following are the strengths of this meta-analysis. First, to decrease the possibility of missing reports, we thoroughly searched five major databases without publication dates or language constrains. Secondly, our analysis followed a predefined protocol to include studies that met rigorous inclusion and exclusion criteria. Thirdly, several stratified analyses were carried out based on several influential study variables, including types of NMSC, types of biological therapy, study quality, treatment years, and sample size. Fourth, all articles included in this study received a relatively high score according to the NOS.2 value identified significant inter-study heterogeneity in the meta-analysis for IBD patients, which was understandable given the wide range of differences across studies regarding recruited participants, treatment drugs, and other study characteristics. Therefore, we conducted subgroup analysis for different research characteristics, partially explaining the heterogeneity among the studies. Also, more studies assessing the association of biological therapy and the incidence of NMSC in patients with IBD and psoriasis are needed.Nevertheless, this meta-analysis had several limitations. The small number of IBD- and psoriasis-specific studies comparing the risk of NMSC between biologic-treated non-biologic treated patients was the major limitation in this meta-analysis. Despite our thorough search, we only found two articles on IBD and three articles on psoriasis qualified for inclusion. Furthermore, the IMoreover, cohort studies have a more significant chance of irreversible bias, mainly confounding, than other types of work. Although all studies corrected for age and gender for NMSC risk, the possibility of bias from unmeasured confounders, which might result in overestimation or underestimating for effect estimate, also remained. For example, the absence of correction for known risk variables for NMSC, including UVR exposure and race/ethnicity, was observed in the studies included in this analysis.Additionally, although no conclusive evidence of publication bias was found based on the Begg’s and Egger’s tests, as we did not search for unpublished articles or other such literature, we cannot entirely rule out the potential of publication bias. Finally, the estimates were based on 12 studies from Europe or the US, while data from other regions, such as Asia and Africa, were inaccessible. Therefore, we need to be cautious in generalizing the findings in this meta-analysis to other regions’ populations.This study found a positive association between biological therapy and the increased development of NMSC in patients with RA and psoriasis but not in patients with IBD compared with patients not receiving biological therapy. Therefore, biological therapy might be avoided in patients with RA or psoriasis who are at high risk of NMSC. The inner cause for the increased risk of NMSC in IBD patients should be further discussed. Considering the significant heterogeneity of IBD in previously published studies, we propose that further large, well-designed studies on this issue are warranted to enhance assurance. The main risk factors for NMSC should also be taken into consideration in future studies.Additional file 1: Table S1. Methodological quality of studies included in the final analysis based on the Newcastle–Ottawa Scale.Additional file 2: Fig. S1. Forest plot for the subgroup analysis of the risk for non-melanoma skin cancer (NMSC) in patients with common inflammatory diseases receiving biologics compared with patients not receiving biologics classified by (a) types of NMSC, (b) types of biological treatments, (c) treatment years, (d) Newcastle–Ottawa Quality Assessment Scale (NOS) score, and (e) sample size. RA: rheumatoid arthritis; IBD: inflammatory bowel disease; TNFI: tumor necrosis factor inhibitor; BCC: basal cell skin cancer; SCC: squamous cell skin cancer; RR, relative risk; CI, confidence interval. The shadow boxes represent point estimates, and the horizontal lines represent 95% CIs. The weight of the research is reflected by the size of the box. Diamonds represent pooled estimates, with their tips representing 95% CIsAdditional file 3. Detailed search strategy.


We evaluate the association of hypertension with PD in an Asian population and performed a meta-analysis on similar studies to address the effect of hypertension on PD risk. A matched case-control study involving 1342 Chinese subjects  was conducted. Hypertension increases PD risk by 1.9 times [OR 1.86 (1.46–2.38)]. The literature search identified 618 studies initially; however, only three matched case-control studies  met the inclusion criteria for meta-analysis. Overall analysis showed that hypertension decreases PD risk by 0.2 times [OR 0.80 (0.66–0.96)]. Hypertension increases PD risk by 1.9 times in our Asian population. However, a meta-analysis comprising of Caucasian populations showed a protective effect of hypertension suggesting that ethnic differences or other genetic or environmental factors may contribute to the divergent observation. Early diagnosis and treatment of hypertension may potentially reduce the risk of PD, at least in our population. PD incidence increases with age, with a rate of 32 per 100 000 among individuals 50 years or above in the Singapore population comprising predominantly of ethnic Han Chinese2. The disease burden is expected to increase in rapidly ageing populations.Parkinson’s disease (PD) is a debilitating neurodegenerative disorder clinically characterized by resting tremor, bradykinesia, rigidity, gait disturbances, and postural instability. In addition to motor impairments, PD is associated with many non-motor features, such as constipation, lower urinary tract symptoms, sleep disturbances, and neuropsychiatric symptoms3. Environmental risk factors that might predispose individuals to PD remain to be fully elucidated. Hypertension is one of the most common chronic diseases globally. In our population, about 20% aged 18−69 years had hypertension4. Hypertension is known to increase the risk of cardiovascular diseases5 and is one of the leading risk factors for cerebral small vessel disease, as demonstrated in human and animal studies7. Hypertension accounts for high risk of mortality and substantial disability8. Dietary and lifestyle changes can potentially improve blood pressure control, preventing hypertension, and decrease the risk of associated medical complications5.The pathophysiology behind PD is complex and likely to be influenced by the interplay of genetic and environmental factors9, with studies showing divergent results11. The neurodegenerative effect is postulated to be caused by vasculopathy leading to strategic infarcts that could affect dopaminergic cells and disrupt neuronal connections12. The vasculopathy pattern differs between Asians and Caucasians with Asians being more likely to suffer from small vessel strokes14. To date, there is a paucity of data on the association between hypertension and PD.The etiologic link between hypertension and PD has been debatedTo address the gaps in current knowledge, we investigate the association between hypertension and PD risk utilizing a matched case-control study. In addition, we performed a pooled analysis of similar matched case-control studies to further explore the association of hypertension on PD risk.A total of 1342 subjects comprising of 671 PD and 671 age and gender-matched controls were included.The mean age of cases and controls were 63.9 ± 9.7 and 63.5 ± 9.8 years respectively. Gender distribution was equal for both cases and controls, with 59.5% males and 40.5% females in each group. PD cases were more likely to have hypertension 46.1%) .A meta-analysis of matched case-control studies showed a protective effect of hypertension on the risk of PD. The association was significant after adjustment for potential confounders [OR 0.80 (0.66–0.96)] Fig. . There wUtilizing a closely matched case-control methodology, our study suggested that hypertension is associated with an increased PD risk in our Asian population. In comparison, pooled analysis of matched case-control studies, which comprised of Caucasian populations showed that hypertension is associated with reduced risk of PD.9. These discrepancies could be explained by heterogeneity in the study population and design. Although subgroup analysis or meta-regression was not performed, we observed that all three included studies had a similar effect, indicating an inverse relationship of hypertension with PD17. This could suggest that the source of heterogeneity might be due to variations in magnitude instead of the direction of effects. The heterogeneity could also arise from different study designs. The studies included had mostly different numbers of PD cases and controls. One study was a nested case-control study, which explained the relatively high sample size of 3637 PD and 3637 controls, compared to the other two case-control studies with varying sample sizes as well. Moreover, the studies have adjusted for different confounders, which further substantiate that the heterogeneity could come from the adjustment for different covariates. There were no Asian studies included in this meta-analysis as these studies did not meet our inclusion criteria. Either no matching was performed, or the focus of analysis was not between PD cases and controls.Epidemiologic studies on hypertension generally had inconsistent associations with PD18. A low prevalence of hypertension was seen in Koreans and Chinese compared with Caucasians20. However, certain Asian races such as Filipinos had a higher prevalence of hypertension20.In our analysis, all three published studies had a higher proportion of hypertension in the controls. Whereas in our study, a higher proportion was observed in PD cases. This divergent observation could be explained by a variance in prevalence of PD and hypertension in different populations, ethnic-specific genetic factors, or exposure to environmental agents or lifestyle habits, or a combination of these factors. Both PD and hypertension are chronic diseases with rising prevalence globally. It has been shown that the prevalence of PD is lower in Asian countries such as China and Japan, compared to a higher prevalence in Western populations11, corroborating our observation. If we have included studies with non-matched samples in our meta-analysis, there is a shift in trend towards higher odds of PD risk in the presence of hypertension (data not shown). Multiple non-matched Western studies have also established that hypertension increases PD risk22. In the Asian population, cohort studies in Taiwan showed an increased risk of PD in the presence of hypertension24, this is also seen in the Korean population25. However, a study in Japan showed that hypertension is associated with a decreased risk of PD26. These differences suggest potential underlying heterogeneity and subgroup variation, which could be attributed to many factors such as ethnic-specific genes or differences in lifestyle habits.Interestingly, some studies, which did not satisfy our meta-analysis inclusion criteria showed that hypertension was significantly associated with increased risk of PD among Asians16. Given that hypertension is one of the major risk factors of cerebral small vessel disease and has shown to contribute to cognitive decline and Alzheimer’s disease6, it is highly plausible that hypertension could result in greater PD risk. Chronic hypertension leading to infarcts at strategic locations such as the basal ganglia could disrupt dopaminergic cells and neuronal connections, subsequently leading to cerebrovascular dysfunction as suggested by pathological and imaging studies28. Furthermore, long-standing hypertension increases the risk of oxidative stress, resulting in reduced cerebral perfusion, which could decrease oxygen delivery to brain regions implicated in PD that are highly susceptible to ischemia30.Emerging studies suggest that hypertension might share similar pathogenic pathways with PD, although exact mechanisms remain to be elucidated and require further investigations from in-vitro and animal models. The advantageous effect of hypertension on PD was postulated to be due to persistent elevation of cerebral blood flow31. PD-induced harmful alterations to the autonomic nervous systems were shown. However, in addition to the sympathetic nervous system, the parasympathetic function is impaired as well. This results in the decreased ability to regulate blood pressure changes leading to a higher ambulatory blood pressure variability in PD32. A greater understanding of the pathogenesis pathways would be necessary to form the basis for the future development of pharmacological interventions in PD.It has been suggested that reduced sympathetic activity in PD, due to decreased in levels or blunted activity of hormones involved in the sympathetic nervous system, could possibly explain the inverse relationship between hypertension and PDOur study involved a closely matched case-control data set. In addition, as Asian ethnicity may be varied (just like Caucasian race), we have only included ethnic Han Chinese, thus reducing racial difference as a confounding factor. It is unclear if our findings could be extrapolated to other populations since the underlying gene and environmental exposures may be ethnic-specific. Further studies in well-designed large prospective studies to better understand the interplay of environmental and gene factors, and the pathogenic mechanisms of PD are warranted. We have included matched studies with at least 200 cases and 200 controls in our meta-analysis because small unmatched studies are at a higher risk of both type 1 and type 2 errors. Future matched case-control studies in the Asian populations should be performed to corroborate our findings.33.There is potential for our findings to contribute to the institution of behavioral modification interventions including lifestyle modifications or dietary recommendations for PD patients and asymptomatic high-risk populations to mitigate disease risk. Studies at the molecular level can be used to investigate the effects of hypertension on PD. These will be essential in elucidating the shared biomolecular mechanisms underlying both chronic diseases. A better understanding of the pathophysiology could form the basis for the future development of individualized disease-modifying therapies in PDWe demonstrated that hypertension increases PD risk by 1.9 times in our Asian population. However, our meta-analysis comprising studies in Caucasian populations was unable to replicate the finding suggesting that ethnic differences may contribute to divergent observation. Future longitudinal studies to evaluate the role of ethnic-specific gene variants in mediating the interaction between hypertension and PD will be of interest. Our study also suggests that early diagnosis and treatment of hypertension may potentially reduce the risk of PD, at least in our population.34 were recruited as cases at two major movement disorders centers. Healthy controls were volunteers without neurodegenerative diseases and examined by the investigators. We have only included ethnic Han Chinese in our study. For every PD patient, we looked for a similar age (± 2 years) and gender-matched control subjects.Patients diagnosed with PD by movement disorder neurologists according to the UK PD Society Brain Bank clinical diagnostic criteriaThe study received approval from the Singapore General Hospital/Singhealth institutional ethics committee (2002/08/A). Subjects have given written informed consent. The methods were carried out in accordance with the approved guidelines.35. A response was coded as “No” if the subject never had hypertension and “Yes” if the subject was diagnosed with the condition by their physicians, supported by the use of prescribed medications. In addition to self-reported blood pressures, objective measurement was performed in clinics during recruitment. Subjects with high blood pressures during the clinical examination were subsequently diagnosed with hypertension and classified as hypertension for the purpose of this study.We used a previously validated questionnaire to obtain clinical information on hypertension, demographics, and family history of movement disorders from participantsA database search using PubMed from the last 20 years (2000–2020) was performed. The following search terms were used: (“Parkinson’s disease” or “Parkinson disease”) and (“hypertension” or “high blood pressure”). Studies that met the following eligibility criteria were included for meta-analysis. Inclusion criteria included: (1) publications limited to English and human subjects only; (2) at least 200 cases and 200 controls comparing hypertension between PD cases and controls; (3) age and gender must be closely matched in cases and controls (age difference of at most ±3 years); (4) available data on the frequency of hypertension, age, and gender in cases and controls; (5) assessment of hypertension was provided; (6) PD diagnosis criteria were provided; (7) available data on odds ratio (OR), relative risk (RR) or hazard ratio (HR) and 95% confidence interval (CI).Studies were excluded based on the following criteria: (1) studies without original data such as reviews or letters; (2) not matched for both age and gender; (3) wide age difference in matching (> ±3 years); (4) OR, RR or HR, and 95% CI were not provided or could not be derived from calculations based on available data.I2 > 50%. Subgroup analysis and meta-regression were not performed given the small number of available studies36. The significance level was set at p < 0.05. An I2 value that is more than 50% would indicate the presence of significant heterogeneity38.Subjects with demographic data for age, gender, and hypertension were included in the analysis. Baseline characteristics were reported using descriptive statistics and were compared between PD and controls using McNemar test. OR with 95% CI were derived from conditional logistic regression models adjusted for age and gender to study the independent effect of hypertension with PD risk. In the meta-analysis, a random-effects model was used to derive pooled OR with 95% CI given Data analysis was performed using SPSS . Meta-analysis was performed usingReview Manager (RevMan) [Computer program]. Version 5.4.1 The Cochrane Collaboration, 2020.Further information on research design is available in the Reporting Summary


Increasing evidence has suggested an association of adiponectin gene polymorphisms rs1501299, rs2241766, rs266729 and rs3774261 with risk of nonalcoholic fatty liver disease (NAFLD). This correlation has been extensively meta-analyzed for the first two polymorphisms, but not the second two.The PubMed, EMBASE, Google Scholar, and China National Knowledge Infrastructure databases were searched for relevant literature. Odds ratios (ORs) and 95% confidence intervals (CIs) were calculated.P < 0.001; recessive, OR 2.35, 95% CI 1.86-2.95, P < 0.001; dominant, OR 1.84, 95% CI 1.34-2.53, P < 0.001; homozygous, OR 2.69, 95% CI 1.84-3.92, P < 0.001; and heterozygous, OR 1.72, 95% CI 1.28-2.32, P < 0.001. This association between rs266729 and NAFLD risk remained significant for all five models among studies with Asian, Chinese and Caucasian samples. The rs2241766 polymorphism was associated with significantly higher NAFLD risk according to the recessive model .A total of 10 case-control studies on rs266729  and 3 case-control studies on rs3774261 (562 cases and 793 controls) were included. Meta-analysis showed that rs266729 was associated with significantly higher NAFLD risk based on the following five models: allelic, OR 1.72, 95% CI 1.34-2.21, Polymorphisms rs266729 and rs3774261 in the adiponectin gene may be risk factors for NAFLD. These findings may pave the way for novel therapeutic strategies, but they should be verified in large, well-designed studies. Nonalcoholic fatty liver disease (NAFLD), also known as metabolism-associated fatty liver disease , is rapiAMP1 gene on human chromosome 3q27, and it consists of three exons and two introns. Genetic and epigenetic changes in the adiponectin gene may reduce adiponectin levels in plasma and dysregulate hepatic lipid metabolism, which may help explain differences in NAFLD risk among individuals  those seven terms in combination with polymorphisms, SNP, variant, variants, variation, genotype, genetic or mutation; and (c) all of the above terms in combination with nonalcoholic fatty liver disease or NAFLD. Only studies involving humans were considered. Reference lists in original and review articles were searched manually to identify additional studies. In the case of multiple studies involving overlapping samples, only the largest study was retained.The PubMed, EMBASE, Google Scholar, Web of Science and China National Knowledge Infrastructure (CNKI) databases were searched up to October 20, 2021 without language restrictions using the following search terms: (a) Studies were included if they met the following criteria: (a) studies had a case-control design to assess the association of adiponectin rs266729 or rs3774261 with NAFLD risk; (b) all patients were diagnosed with NAFLD based on the following diagnostic criteria: abnormal levels of aspartate aminotransferase and alanine aminotransferase persisting for at least 6 months, or evidence of fatty liver based on ultrasonography and/or evidence of diffuse fatty liver based on other imaging examinations, or liver histology; (c) the full text was available and it reported genotype frequencies in cases and controls, or sufficient data to estimate odds ratios (ORs) and 95% confidence intervals (CIs).Studies were excluded if they: (a) were not a case-control study; (b) did not report precise genotypes; (c) were duplicate publications of data from the same study; (d) were meta-analyses, letters, reviews, or editorial articles; (e) investigated other polymorphisms of adiponectin.P value for Hardy-Weinberg equilibrium among controls, and matched parameters.Two authors (YTZ and LYL) independently selected eligible studies and extracted the following data: first author’s name, year of publication, ethnicity, country, sample size, type of controls, genotyping method, genotype distribution, The quality of included studies was assessed independently by two investigators (YTZ and LYL) using the Newcastle–Ottawa Scale . Scores The strength of association of rs266729 and rs3774261 with NAFLD risk was calculated in terms of unadjusted ORs with 95% CIs based on genotype frequencies in cases and controls. The significance of pooled ORs was determined using the Z test, with P < 0.05 defined as significant. Meta-analysis was conducted using a fixed-effect model when P > 0.10 for the Q test, indicating lack of heterogeneity among studies; otherwise, meta-analysis was conducted using a random-effect model. All statistical tests for meta-analyses were performed using Review Manager 5.3 (Cochrane Collaboration). Publication bias was assessed using Begg’s funnel plot and Egger’s weighted regression in Stata 12.0 , with P < 0.05 considered statistically significant.The search strategy retrieved 313 potentially relevant studies, 277 of which were excluded on the basis of titles and abstracts Figure 1Ten studies –45 focusP < 0.001; recessive, OR 2.35, 95% CI 1.86-2.95, P < 0.001; dominant, OR 1.84, 95% CI 1.34-2.53, P < 0.001; homozygous, OR 2.69, 95% CI 1.84-3.92, P < 0.001; and heterozygous, OR 1.72, 95% CI 1.28-2.32, P < 0.001 . But this association could not be found in the Chinese population .Meta-analysis of three studies –47 involpulation Table 2P value associated with Hardy-Weinberg equilibrium was less than 0.05  , 58 and P value associated with Hardy-Weinberg equilibrium was < 0.001 in three studies (Our results should be interpreted with caution in light of several limitations. First, the controls in one study  had diab studies , 42, 47,The available evidence suggests that SNPs rs266729 and rs3774261 in the adiponectin gene are risk factors for NAFLD. If our results can be verified in large, well-designed studies, they may help pave the way for novel therapeutic strategies.The original contributions presented in the study are included in the article/Designed the study: L-YL and Y-TZ. Searched databases and collected full-text papers: T-MX and C-XW. Extracted and analyzed the data: L-YL and Y-TZ. Statistical analyses: J-YC. Wrote the manuscript: Y-TZ. All authors reviewed the manuscript. All authors contributed to the article and approved the submitted version.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.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.


Gallstone disease (GSD) is a common and costly biliary disorder. Multiple studies have investigated the associations between blood lipid metabolism and GSD risk; however, the results are inconsistent. This research aimed to comprehensively evaluate the relationships among serum total cholesterol, low-density lipoprotein (LDL) cholesterol, high-density lipoprotein (HDL) cholesterol, triglycerides and GSD risk.Firstly, a multi-center cross-sectional study was carried out. Subjects who participated in the health examination in three hospitals between January 2015 and May 2020 were recruited. Multivariable logistic regression was used to investigate blood lipid metabolism associated with GSD risk. Then, a meta-analysis was performed to verify the associations further. Medline and Embase databases were systematically searched before June 10, 2021. The DerSimonian and Laird random-effect model was utilized when the heterogeneity was high; otherwise, fixed-effect model was adopted.P = 5.97 × 10− 16 in high vs low model; OR = 0.974, P = 6.07 × 10− 05 in per unit model] and triglyceride  were related to GSD risk in the two models.There were 548,934 eligible participants included in the multi-center study, and 45,392 of them were diagnosed with GSD. The results demonstrated that total cholesterol and HDL cholesterol were negatively associated with GSD risk in both high vs. low model and per mmol/L increase model, while triglyceride was positively associated with GSD risk in the per unit increase model. In the meta-analysis, 104 studies with approximately 3 million participants were finally included. The results verified that HDL cholesterol [odds ratio (OR) = 0.636, The findings indicated that low HDL cholesterol levels and high triglyceride levels were risk factors for GSD. This study provides a basis for identifying the population at high risk for GSD and implementing tertiary prevention strategies for GSD, thus contributing to GSD prevention as well as disease burden relief.The online version contains supplementary material available at 10.1186/s12944-022-01635-9. Gallstone disease (GSD) is one of the most common biliary diseases, afflicting about 10–20% of the population in European countries . In the Most gallstones are composed of cholesterol, and hence the role of cholesterol metabolism in the mechanism of gallstone formation has long been a research focus in the etiology of GSD. Several studies have evaluated the relationship between serum lipids levels and GSD risk , 8. HoweIn this research, a multi-center cross-sectional study in China was firstly conducted to estimate the associations between serum lipids  and GSD risk. Additionally, meta-analysis was conducted to further verify these associations.This cross-sectional study was performed in compliance with the guidelines of the Strengthening the Reporting of Observational Studies in Epidemiology (STROBE) statement .Participants in this multi-center cross-sectional study were recruited from the physical examination centers between January 2015 and May 2020 in three hospitals in China. The three hospitals are the First Affiliated Hospital of Chongqing Medical University Jinshan Hospital, the People’s Hospital of Kaizhou District in Chongqing city, and Tianjin Medical University Cancer Institute and Hospital. The participants who met the following criteria were recruited: (i) underwent ultrasonography examination; (ii) had complete demographic, anthropometric and biochemical indexes, including age, gender, height, weight, waist circumference, blood glucose, diastolic/systolic blood pressure (DBP/SBP), and indicators for liver function and kidney function. If the subject participated in multiple health examinations, the latest data were selected. This research was approved by the ethics committee of West China Fourth Hospital and West China School of Public Health, Sichuan University (Gwll2021055), and was conducted according to the ethical guidelines of the 1964 Declaration of Helsinki and its later amendments.The fasting blood samples were collected from the subjects, and then measured in laboratories within one hour. The biochemical analyzers were utilized to estimate the serum concentrations of HDL cholesterol, LDL cholesterol, TC, triglyceride, fasting blood glucose (FBG), total bilirubin (T-bil), alanine transaminase (ALT), aspartate transaminase (AST), creatinine (Cr), uric acid (UA), and urea nitrogen (UN). All biochemical indexes were measured by following the international standard protocol independently in each hospital.Ultrasonography was performed by experienced radiologists. The diagnosis of GSD was based on one of the following two criteria or both: (i) one or more hyperechoic structures in the gallbladder or biliary system, which were acoustic shadowing or gravity-dependent; (ii) no sight of the gallbladder in patients who underwent cholecystectomy due to gallstones. In this cross-sectional study, the lipid markers were classified into three groups according to the Chinese adult dyslipidemia prevention guide (2017 edition), TC (mmol/L): > 5.7, 3.1–5.7, and < 3.1; triglycerides (mmol/L): > 1.7, 0.4–1.7, and < 0.4; LDL cholesterol (mmol/L): > 3.1, 2.07–3.1, and < 2.07; and HDL cholesterol (mmol/L): > 2.0, 0.9–2.0, and < 0.9.The current meta-analysis was conducted according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) statement  and the In this meta-analysis, two corresponding authors (X.S. and X.W.) developed the search strategy, inclusion criteria, and exclusion criteria. Two authors (M.Z. and M.M.) independently conducted literature searching, article selection, data extraction, and quality evaluation. All inconsistent data were discussed and resolved by the corresponding authors. The PubMed and Embase databases were searched to screen the relevant researches published in English before June 10, 2021. The exact keywords are listed in Additional file Publications were eligible to be included if: (i) they were observational studies conducted in humans, including cross-sectional, case-control, and cohort studies; (ii) the primary outcomes were risk of GSD and/or mean difference in lipid levels between GSD patients and healthy individuals; (iii) they directly provided risk estimates such as hazard ratio (HR), relative risk (RR) and odds ratio (OR) with 95% confidence intervals (CIs), coefficient β and standard error (SE), or provided enough data to calculate these risk estimates; or (iv) they provided mean and standard deviation (SD) in both cases and controls. When the same population was used in multiple research papers, the most recent study or the one with the largest sample size were selected. Irrelevant studies and relevant literature reviews with insufficient data were excluded.P-value , and mean levels of lipids and SD in GSD and non-GSD groups.Data were extracted from a predesigned sheet containing the PubMed Unique Identifier (PMID), first author, publishing year, exposure factor, source of population, geographic background, study period, study design, sample size of cases and controls, matched information, number of male and female participants, effects of associations and the corresponding 95%CIs, coefficient β, SE or Risk of bias and quality assessment were conducted for the included cohort and case-control studies by the Newcastle-Ottawa scale (NOS), and the Agency for Healthcare Research and Quality (AHRQ) for cross-sectional studies as suggested , 13. NOSt-test was utilized to evaluate the difference between GSD group and non-GSD group. The categorical variable was expressed as number and percentage, and was compared by the Chi-square test. Multivariate logistic regression models were applied to evaluate the associations between serum lipids profile and GSD risk in each hospital. The effects of higher levels and each mmol/L increase in blood lipid levels on GSD were assessed. Finally, the results of the associations between serum lipid levels and GSD risk from the three hospitals were pooled. Subgroup analyses were conducted based on the subtype of GSD, age, and gender.All statistical analyses in the cross-sectional study were performed with SPSS software, version 19 . The continuous variable was presented as mean ± SD, and Student’s Q test was performed to examine the heterogeneity between publications, and the I2 statistic was applied for quantification of heterogeneity. DerSimonian and Laird random-effect model was utilized when the heterogeneity was evident (I2 ≥ 50%); otherwise, fixed-effect model was adopted. For the included studies that provide the mean levels of blood lipids and SD, meta-analysis for continuous variables was conducted to obtain standard mean difference (SMD). For the studies that provide the estimates of lipid levels on GSD with OR/HR/RR and 95%CI, meta-analyses were performed on the effects of higher lipid levels or each increasing unit of blood lipid levels on GSD to obtain the pooled ORs and 95%CI. Next, a dose-response meta-analysis was conducted by robust error meta-regression (REMR) model using valid data [P-value< 0.10 was deemed significant as recommended. The significance level of a two-tailed test was set at P-value less than 0.05.The STATA software  was utilized for meta-analyses. Cochran’s lid data , 15. ThrP < 0.05).Overall, 45,392 GSD cases and 503,542 controls were recruited from three hospitals in this multi-center cross-sectional study. The clinical, anthropometric characteristics and laboratory test results of the subjects are presented in Table Multivariate logistic regression analysis revealed that the higher levels of HDL cholesterol, LDL cholesterol and TC were negatively related to GSD risk in the two hospitals in Chongqing Table . The higThen, the results of each hospital were pooled. As shown in Table When cholecystectomy and gallstones were analyzed separately, the higher concentrations of TC and LDL cholesterol were negatively related to cholecystectomy, but LDL cholesterol was related to a high susceptibility to gallstones formation. The high level of TC was associated with GSD risk only in the older population when stratified by age. Additionally, a higher level of LDL cholesterol was positively correlated with GSD risk in younger population, while negatively correlated with GSD risk in middle-aged and elderly populations Fig. a. One mmn = 57), followed by European countries (n = 21), American countries (n = 18), Oceania country (n = 1), and other countries (n = 4). Among these studies, 41.6% (n = 42) were regarded as high quality and 58.4% (n = 59) were moderate quality according to the statement of NOS and AHRQ  also indicated that the associations were generally consistent when stratified by gender, study design, the grade of the quality and geographic background. The heterogeneity was significant in most of the meta-analysis models, and meta-regression showed that the study design, gender, geographic background, and quality of the included studies might be the sources of heterogeneity . Egger’s test indicated eight associations with small-study bias  was observed , triglycerides , LDL cholesterol  and GSD risk were detected.As to the dose-response meta-analysis, a total of 17, 18, 12 and 15 studies reporting on the associations between GSD risk and levels of TC, triglycerides, LDL cholesterol, HDL cholesterol were included, respectively, according to the criteria of the REMR model. The doses ranged from 0 to 15.22, 0 to 4.47, 0 to 7.41, and 0 to 2.85 for TC, triglycerides, LDL cholesterol, and HDL cholesterol, respectively. A linear relationship between HDL cholesterol level and GSD risk  genes are significantly related to higher concentrations of triglyceride and lower concentrations of HDL cholesterol [ABCG5/G8 genes are associated with the risk of GSD [ABCG5/G8 gene products function as a half-transporter to promote cholesterol transport into bile. In patients with GSD, the expression of ABCG5/G8 was upregulated and could affect the cholesterol supersaturation of bile [CYP7A1) gene also result in premature GSD and hypertriglyceridemia [CYP7A1 gene catalyzes the initial steps of cholesterol catabolism and cholic acid synthesis.Genetic factors may also contribute to the relationship between blood lipids and GSD risk. It has been suggested that Q6404 and D19H polymorphisms in the ATP binding cassette protein G5/G8  total cholesterol; (B) LDL-C; (C) Triglycerides; (D) HDL-C.