Meanwhile, in the subgroup analysis, the effect value of fetuin-A in the NASH group was significantly higher than that in the NAFL group (p = 0.036). The findings of this study suggest that elevated fetuin-A and fetuin-B may independently indicate the occurrence of NAFLD. Nevertheless, further research is needed to confirm these results.The magnitude of the effect of fetuin-A and fetuin-B on non-alcoholic fatty liver disease (NAFLD) remains undefined. Therefore, the aim of this study was to synthesize previous findings to obtain a reliable estimation of this relationship. This study was registered in PROSPERO with the number CRD42019126314. Studies published not later than March 2019, examining the relationship between fetuin-A, fetuin-B, and NAFLD, were identified by a systematic search in the electronic databases of the Web of Science, PubMed, Embase, and Cochrane Library. Pooled estimates of standardized mean difference (SMD), calculated using the random-effects model in a meta-analysis, were applied to estimate the strength of the association between fetuin-A, fetuin-B, and NAFLD. Thirty publications were identified and analyzed based on specified inclusion criteria. Collectively, they consisted of 3800 NAFLD participants and 3614 controls. Compared with the controls, significant higher values of the fetuin-A (SMD = 0.83, 95% CI: 0.59 to 1.07, Z = 6.82,  Non-alcoholic fatty liver disease (NAFLD) is now highly prevalent and often related to complications such as liver failure, cirrhosis, and liver cancer. Thus, it has become a major health concern and new worldwide epidemic. Basically, NAFLD is a chronic hepatic steatosis from non-alcoholic fatty liver (NAFL) to non-alcoholic steatohepatitis (NASH) . HoweverTo sum up, there is clear evidence that fetuin-A and fetuin-B are complex metabolic phase reactants that have controversial roles in NAFLD. Therefore, the aim of this study was to synthesize previous findings to obtain a reliable estimation of the relationship between fetuin-A, fetuin-B, and NAFLD.This study was carried out using the guidelines provided in the Cochrane handbook (version 5.1.0) for systematic reviews and meta-analysis, whereas the results were reported according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) . In addiStudies published between January 1959 and March 2019, examining the relationship between fetuin-A, fetuin-B, and NAFLD, were identified by a systematic search in the electronic databases of the Web of Science, PubMed, Embase, and Cochrane Library. Detailed search strategies were designed with help from experienced librarians and are shown in the appendix of the  HS Glycoprotein[Title/Abstract] OR alpha-2HS Glycoprotein[Title/Abstract] OR alpha 2HS Glycoprotein[Title/Abstract] OR alpha2HS Glycoprotein[Title/Abstract] OR Fetuin-A[Title/Abstract] OR Fetuin-A[Title/Abstract] OR AHSG Protein[Title/Abstract] OR alpha2 Heremans-Schmid Glycoprotein[Title/Abstract] OR alpha-2-HS-Glycoprotein[Title/Abstract] OR fetuins[Title/Abstract] OR Fetuin[Title/Abstract] OR Fetuin-B[Title/Abstract] OR Fetuin-B[Title/Abstract])) AND .Titles, abstracts, and full texts of the studies yielded after searching in the electronic databases were screened to meet the following inclusion criteria: (1) Studies were case-control studies; (2) studies reported a criteria used to diagnose NAFLD; (3) studies provided mean and standard deviation (SD) of fetuin-A and fetuin-B, or these could be obtained from respective authors on request; (4) studies were peer-reviewed before publication; and (5) studies were published in English. On the contrary, the exclusion criteria were: (1) Studies were case reports or reviews; (2) studies reported NAFLD in combination with other diseases; (3) studies had fetuin-A and fetuin-B pharmacologically challenged before their measurements; (4) studies did not examine humans; and (5) studies were grey literature (unpublished literature). Two researchers [KA and AK] independently identified the eligible studies, and their discrepancies were resolved by involving the third researcher [JL]. Data related to the following variables were extracted: (1) The first author’s name and year of publication; (2) subject’s characteristics such as Body Mass Index (BMI), mean age , and gender; (3) region of study; (4) participant’s laboratory characteristics such as aspartate aminotransferase (AST), alanine aminotransferase (ALT), glutamyl transpeptidase (GGT), low-density lipoprotein (LDL), high-density lipoprotein (HDL), triglycerides (TG), and insulin resistance index (HOMA-IR); (5) sample characteristics such as the material of sample, and mean and SD of fetuin-A and fetuin-B; and (6) fetuin-A and fetuin-B sample test methods, and storage temperatures. Two researchers [KA and AK] conducted data extraction independently using EpiData 3.0 and Microsoft Excel 2007. Any discrepancies between them were resolved by consensus.The non-randomized Newcastle-Ottawa Scale (NOS) was used to assess the quality of this study . It is a2 statistic [2 statistic was lower than 25%, moderate if it was 25%–75%, and high when it was greater than 75%. Additionally, the overall results were divided into the subgroups, including NASH patients and the control group, and NAFL patients and the control group. In addition, subgroup analysis was conducted to explore whether samples, storage methods, or ethnicity had any influence on the results. However, the subgroup analysis of fetuin-B was not carried out due to inadequate information related to fetuin-B. Furthermore, meta-regression analyses were conducted to explore the source of heterogeneity when this was significantly high. Subjects’ characteristics such as BMI and mean age ; and participant’s laboratory characteristics such as AST, ALT, GGT, LDL, HDL, and TG were taken into account when exploring the source of heterogeneity. Additionally, a sensitivity analysis was performed by repeating the meta-analysis when each was omitted in turn. Publication bias was assessed by examining the symmetry of funnel plots when the number of studies reporting the primary outcomes was at least 10, and the significance of the symmetry was tested using the Egger’s linear regression test [p-values less than 0.05 indicated significant results for all statistical tests.The random effects model was used to pool the estimates under discussion because of the varying populations and criteria used to define outcomes across the eligible studies. This model, therefore, was used to merge standardized mean differences (SMDs) and corresponding 95% confidence intervals (CIs), to compare the concentrations of fetuin-A and fetuin-B between NAFLD patients and the controls, which was calculated as Cohen’s d . The pootatistic . Similarion test . All anaCharacteristics of the 30 eligible studies are presented in p < 0.001), and heterogeneity was considerable (I2 = 91.1%). In addition, subgroup analysis found that fetuin-A concentrations were significantly higher in the NAFL patients than in the controls, with high effect ; and in the NASH patients than in the controls, with high effect , but both with high heterogeneity . Moreover, the effect value, SMD, was significantly higher in the NASH group than in the NAFL group (p = 0.036). However, subgroup analysis showed no difference in the effect values of fetuin-A for plasma and serum samples (p = 0.282), and with respect to RACE (p = 0.155). Nonetheless, the effect values of fetuin-A in relation to storage methods were significantly different, and the effect values of fetuin-A according to frozen storage temperature were higher (p = 0.045) (p = 0.023), and moderate heterogeneity (I2 = 62.3%).= 0.045) . On the p = 0.158), age , AST , ALT , GGT , and TG . However, subsequent meta-regression analysis revealed a negative effect of LDL  and HDL  on SMD. Additionally, the results of meta-regression analysis  of fetuin-A and fetuin-B are shown in p = 0.747), age , ALT , AST , GGT , LDL  and HDL . However, subsequent meta-regression analysis revealed a positive effect of TG on SMD .Nonetheless, as regards fetuin-B and NAFLD, the following clinical characteristics had a negative effect on SMD: BMI , implying that there may be publication bias for this comparison.Sensitivity analysis showed that there was a small change in SMD and the corresponding 95% CI when each study was removed in turn, indicating that the current overall meta-analysis data were relatively stable. Publication bias was not reported for fetuin-B since the number of fetuin-B studies was less than 10. However, there was lack of symmetry in the shape of the Egger’s funnel plot for fetuin-A, and the Egger’s test was significant , thus supporting the previous hypothesis that there are raised levels of fetuin-A in NAFLD. In addition, subgroup analysis showed that the effect value of fetuin-A concentrations was significantly higher on the NASH patients than on the NAFL patients, which may indicate that fetuin-A played an important role in the pathophysiological process of NAFL to NASH. This inference could not be made on fetuin-B due to a lack of relevant studies on the relationship between fetuin-B concentrations and NASH. Therefore, the changes of fetuin-B in NASH should be studied accordingly in the future. However, subgroup analysis showed no difference in the effect values of fetuin-A for plasma and serum samples, and with respect to RACE, whereas the effect values of fetuin-A in relation to storage methods were significantly different, and the effect values of fetuin-A according to frozen storage temperature were higher. Thus, these findings suggest that the effects of fetuin-A were stable in relation to sample sources, storage methods, and race. However, future research should try to use cold chain to transport and store fetuin-A. Noteworthy, similar results were found between fetuin-A concentrations and other diseases. For example, fetuin-A levels were increased in other glucose and lipid metabolism disorders such as type 2 diabetes mellitus and metabolic syndrome. Although NAFLD and these other diseases are similarly identified as chronic diseases, the role of fetuin-A in NAFLD is likely to be more complex.Notwithstanding, a recent study showed some hypothetical mechanisms about the relationship between fetuin-A and NAFLD. Specifically, that study suggested that a high-fat diet would bring excessive free fatty acids and cause hepatocytes and adipocytes to secrete fetuin-A, which can send chemical attraction signals and induce macrophages to infiltrate into adipose tissue . In thisIn addition, it has been confirmed from extensive studies of genetic epidemiology that polymorphisms in the gene encoding fetuin-A, the alpha-2-Heremans-Schmid glycoprotein (AHSG) gene, may affect glucose and lipid metabolism disorders’ pathophysiology progression in some higher-risk groups. Furthermore, metabolic syndrome and type 2 diabetes mellitus are concerned with the single nucleotide polymorphisms (SNPs) of two risk alleles in AHSG (rs2518136 and rs4917) . TherefoIn addition, clinical and experimental studies concurred with the hypothesis of a relationship between NAFLD and fetuin-A. Besides, it is well known that fetuin-A is significantly increased in hepatocellular injury and mediates insulin resistance as well as impairs glucose tolerance . MoreoveAlthough previous studies have shown that fetuin-A and fetuin-B may be affected by other clinical parameters in NAFLD, this study has shown that fetuin-A might be a potential predictor of the occurrence of NAFLD, independent of age, AST, BMI, ALT, GGT, TG, LDL, and HDL. However, HDL and LDL had no significant influence on the effect size of fetuin-A in NAFLD.Considering that increased activity of GGT, AST, and ALT generally indicates chronic and progressive hepatitis lesions, it is no wonder that GGT, AST, and ALT showed a positive influence on NAFLD . As regaDespite the reasons for the negative effects of the levels of LDL and HDL found in this study, not being well known, it is common for LDL and HDL cholesterols to play a counterproductive role and a protective role, respectively, in NAFLD and other metabolic diseases. Thus, it is likely that LDL levels might be affected by fetuin-A. This phenomenon is relevant to the impairment of SIRT1 and AMPK energy sensors, in which fetuin-A may play an intermediary role in lipid-induced hepatic fat lesions . FurtherHowever, unlike for fetuin-A, there was a negative linear relationship between fetuin-B and age, AST, BMI, ALT, GGT, HDL, and LDL, but a positive linear relationship between fetuin-B and TG. Even though these results seem beyond our understanding, previous studies have shown that there was a great positive linear relationship between the natural progression of fetuin-B in NAFLD and hepatic steatosis. Therefore, this may explain why the relationship between these indicators and fetuin-B was different from that between these indicators and fetuin-A. Although there may be a negative correlation between fetuin-B and liver fibrosis, this study did not consider exploring sources of heterogeneity with respect to liver fibrosis because the liver fibrosis scores of the patients varied considerably across the eligible studies. Thus, the relationship between fetuin-B and liver fibrosis needs to be further studied . In addiThis meta-analysis has some limitations. First, these findings were obtained from case-control studies, which cannot make causal inferences. Therefore, prospective studies should be considered in the future to explore the relationship between fetuin-A, fetuin-B, and NAFLD at different stages. Second, confounding factors that may affect fetuin-A and fetuin-B concentrations, such as race and cigarette smoking, were not analyzed in this study because these were not reported in the eligible studies. Third, studies on fetuin-A and fetuin-B concentrations between NAFLD patients and the controls showed high heterogeneity, which is not surprising given the large differences with respect to the study area and study population. Therefore, future original studies should consider making these investigations on homogeneous samples.To sum up, the findings of this study showed that there were significantly higher fetuin-A and fetuin-B concentrations in NAFLD patients than in the controls. Thus, fetuin-A and fetuin-B might be potential predictors of the occurrence of NAFLD. Notably, fetuin-A may play an important role in the pathophysiological process of NAFL to NASH. These findings, therefore, provide a rationale for evaluating fetuin-A and fetuin-B in the pathophysiological process of NAFLD, and might open up new perspectives in early diagnosis, identification of novel biomarkers, and providing novel targets for pharmacological interventions.


Summary odds ratios (ORs) and 95% confidence intervals (CIs) were used to assess the strength of association of SNPs in FcεRIβ gene with allergic diseases risk. They found that FcεRIβ E237G  and −109 C/T  were risk factors for allergic diseases. Guo et al.’s findings are interesting, but we found that several issues should be clarified after carefully reading the paper. Here, we intended to comment on these data clarifications.Guo  Dear editor,FcεRIβ) polymorphisms and allergic diseases risk in Medline, Embase, Web of Science, Chinese National Knowledge Infrastructure, and Wanfang databases. No limit of start year and month was set, and the updated time was August 2019. The terms, search strategies, and inclusion/exclusion criteria were the same as reported by Guo et al. [et al.’s paper [We researched the relevant studies about the association between the high-affinity IgE receptor β chain  risk. The statistical analysis methods and software used in this comment were the same as reported by Guo et al., unless otherwise indicated [Considering the above-listed mistakes or errors in Guo reliable . To overndicated .P-values for Hardy–Weinberg Equilibrium (HWE) test, were shown in The main characteristics of the eligible studies ,20,22–42FcεRIβ C-109T polymorphism with allergic diseases risk. Overall, no significant associations between C-109T polymorphism and allergic diseases risk were observed : 0.909–1.102 for CC+CT vs. TT and OR = 1.015, 95% CI: 0.788–1.307 for CC vs. CT+TT, respectively). When subgroup analyses by ethnicity (Asian and Caucasian), allergic disease classification  and HWE (in and not) were performed, we did not find any statistically significant associations of C-108T polymorphism with allergic diseases risk  (ctively) . When rectively) . When stctively) . No signctively) . In subgctively) .et al.’s [et al.’s [et al. [et al.’s result in White population (OR = 3.693). When merging the result of study 1 with study 2, the overall OR significantly decreased. And when summarizing the results of the first three studies  the overall OR looked bigger than the second. As a result, the overall ORs would generate a V-form change in the forest plot of cumulative meta-analysis. The alarming changes in the plot of cumulative meta-analysis also indicated that there was high heterogeneity of the results of the included individual studies.We also performed a cumulative meta-analysis which accumulated the evidence about association of E237G variant with allergic diseases risk in the order of publication year of individual study. We observed that the association of EG/GG genotypes with increased allergic diseases started to become significant for the first time when Zeng et al.’s  study puet al.’s  study puSensitivity analysis was performed by sequentially omitting each individual study in the order of publication year and the pooled ORs were estimated repeatedly, which was used to evaluate the stability of the results of present meta-analysis. The sensitivity analysis showed that the association of EG and GG genotypes with increased risk of allergic diseases maintained statistically significant when removing any each individual study . Egger’s. some unpublished negative results studies thus could not be included in this analyses might result in an over-estimated association of E237G with allergic disease risk.There are some inherent limitations of meta-analysis which should be taken into consideration when using the results of this comment. First, there was high heterogeneity in this meta-analysis, especially in the case of association of E237G variant with allergic diseases risk. Although, subgroup analyses were performed on the basis of ethnicity, allergic disease category and HWE, heterogeneity among the included studies still be statistically significant in all subgroups. Second, publication bias tests indicated that the probable existence of publication bias, i.e.’s study [In conclusion, the results of Guo et al’s study  should b’s study .


To resolve the incongruent outcomes yielded from different single studies, we conducted the most up-to-date meta-analysis of IL4 gene −589C/T (rs2243250) polymorphism and susceptibility to asthma.Numerous investigations have previously evaluated the association of A systematic literature search was performed in ISI web of science, Scopus, Medline/PubMed databases prior to September 2020, and the pooled odds ratio (OR) and their corresponding 95% CI were calculated to determine the association strength.IL4 gene −589C/T polymorphism increased the risk of asthma across all genetic models, including dominant model (OR = 1.22), recessive model (OR = 1.17), allelic model (OR = 1.21), and TT vs. CC model (OR = 1.34), but not the CT vs. TT model. The subgroup analysis by age indicated that IL4 gene -589C/T polymorphism was significantly associated with asthma risk in both pediatrics and adults. Additionally, the subgroup analysis by ethnicity revealed significant association in Asian, American, and Europeans. Finally, subgroup analysis by East Asian and non-East Asian populations indicated significant associations.Literature search led to retrieving of 49 publications (55 case-control studies) containing 9572 cases and 9881 controls. It was revealed that IL4 gene -589C/T polymorphism was a susceptibility risk in both pediatrics and adults in the whole and different ethnic groups.The current meta-analysis revealed that  Asthma is one of the most common atopic disorders of the respiratory tract, which results in wheezing, coughing, breathlessness, and bronchial obstruction . The preIL4 gene polymorphisms, like promoter region (C + 33 T) SNP  OR “asthmatic”) AND (“interleukin-4” OR “IL-4” OR “rs2243250”) AND (“single nucleotide polymorphism” OR “SNP” OR “polymorphisms” OR “mutation” OR “variation”). No restrictions were placed on language, sample size, population or publication date.The retrieved publications by primary literature search were imported into Endnote X8 software. The duplicate studies were removed and title and abstract of remain studies were reviewed by two investigators. Full-text verification was performed if we could not categorize studies based on title and abstract. Any disagreements during study selection was discussed and resolved by consensus.The following inclusion criteria were used to distinguish eligible studies: i) studies with distinct case and control group evaluating the association between IL-4 C589T polymorphism and susceptibility to asthma; ii) studies with calculable or extractable data for odds ratio (OR) and 95% confidence intervals (CIs); iii) studies with sufficient data for alleles and genotypes in case and control group. The duplicates, reviews, book chapters, and meta-analysis were excluded. The application of these criteria results in 49 qualified studies for the meta-analysis.Two of our authors independently and according to an extraction checklist extracted the following data: the first author, journal and year of publication, country of origin, ethnicity, number of subjects in the case and the control groups for each gender, mean or range of age, genotyping method, genotype counts in the case and the control group. The quality of each study was assessed using the Newcastle-Ottawa Scale (NOS) criteria . StudiesP value< 0.1 considered statistically significant) and I2-test . In the presence of heterogeneity random effect model (REM) was used, however fixed effect model (FEM) was applied in homogeneous condition [P value< 0.05 considered statistically significant) [Statistical analyses were carried out using STATA  and SPSS . The strength of association between polymorphism and asthma susceptibility was estimated by odd ratios (ORs) and 95% confidence intervals (CIs) for the dominant model, recessive model, allele contrasts, and additive comparison. Heterogeneity among included studies was measured via Q statistics  , 25. Addn = 301), 1820 articles remain for abstract and full-text screening. Of 1820 articles, 1612 were excluded base on title and abstract and 159 articles based on full-text reading. Ultimately 49 publications with 9579 cases and 9881 controls met the inclusion criteria and their data were extracted for meta-analysis. Among these 49 publications, four of them, including Basehore et al. [Our primary search retrieved 2121 potential articles. After removing of duplicate articles , recessive model , allelic model , and TT vs. CC model , except CT vs. TT model  , adults (21 articles) and mixed . The results highlighted a predisposing role of up Table , Fig. 3.IL4 gene -589C/T SNP in Asian population increased susceptibility of asthma across all genotype models except dominant model  and CT vs. CC model . Moreover, in contrast with effect of IL4 gene -589C/T SNP on the risk of asthma in American populations, a significant positive association was detected in European population thorough dominant model , allelic model , TT vs. CC model , and CT vs.CC model  (Table In this subgroup we categorized studies by their continent: including Asia (32 articles), Europe (11 articles), America (10 articles), Africa (1 article), and Oceania (1 article). Since there was only one study for each one of the African and Australian population, these studies were excluded from the analysis. The results indicated that presence of IL4 gene -589C/T polymorphism and the risk of asthma across in all genotype models of East Asians and three genotype models of non-East Asian including; recessive model , allelic model , TT vs. CC model   and non-East Asian (35 articles) title revealed the significant association between IL4 gene -589C/T SNP and asthma risk in Arab population. Also, except recessive model  other genotype models in African-American population were significant including dominant model , allelic model , TT vs. CC model , and CT vs. CC model . Conversely, all genotype models were significant in Caucasians and presence of IL4 gene -589C/T SNP increase risk of asthma , African-Americans (9 articles), and Arabs (5 articles) showed that there was no significant association between Meta-regression analyses were performed to explore potential sources of heterogeneity among included studies Table . The finIL4 gene -589C/T polymorphism  polymorphism has been widely investigated in susceptibility to asthma. It has been shown that the T allele of this SNP leads to increased affinity of the binding of transcription factors in comparison to the C allele, leading to overexpression of IL4 mRNA [IL4 gene −589C/T SNP impresses the IL-4 expression and, hence, could affect the asthma susceptibility.Asthma is a complicated pulmonary disease, characterized by airway hyperresponsiveness, airway inflammation, and airway remodeling , 76. DurIL4 mRNA , 80. As IL4 gene −589C/T SNP with the risk of asthma. Wang et al. in 2012 indicated that the T allele of IL4 gene −589C/T SNP increased the risk of asthma (OR = 1.12). Basically, individuals carrying the T allele had a 24% increased risk of asthma in comparison to the CC homozygote model. Subgroup analysis revealed the association of this polymorphism in the Caucasians [Previously, three meta-analysis studies have attempted to disclose the association of ucasians . In addiucasians . This meucasians  by evaluIL4 gene -589C/T polymorphism increased the risk of asthma across all genetic models, including dominant model (OR = 1.22), recessive model (OR = 1.17), allelic model (OR = 1.21), and TT vs. CC model (OR = 1.34), but not the CT vs. TT model. Furthermore, subgroup analysis by age indicated that IL4 gene -589C/T polymorphism was significantly associated with asthma risk in both pediatrics and adults. The subgroup analysis by ethnicity revealed significant association in Asian, American, and Europeans. Finally, subgroup analysis by East Asian and non-East Asian populations indicated significant associations.Our analysis was performed on 55 case-control studies containing 9572 cases and 9881 controls. It was observed that IL4 gene -589C/T polymorphism association with asthma susceptibility, regardless of the effect of confounding factors, like age, sex, environmental factors, and contribution of other genes in LD with IL4 gene. Second, we did not analyze other genes that could be contributing in understanding of cytokine involvement in the susceptibility to asthma.This meta-analysis bears some limitations and caveats. First, the analysis was according to crude estimation of IL4 gene 589C/T polymorphism and asthma risk prior to September 2020. Our meta-analysis further confirmed some results of the previously performed meta-analysis, while rejected some of them. In a whole, IL4 gene -589C/T polymorphism increased the risk of asthma across all genetic models. Moreover, the subgroup analysis by age indicated that IL4 gene -589C/T polymorphism was significantly associated with asthma risk in both pediatrics and adults. Also, the subgroup analysis by ethnicity revealed significant association in Asian, American, and Europeans. Ultimately, subgroup analysis by East Asian and non-East Asian populations indicated significant associations.All in all, here we carried out the most up-to-date analysis of the


We tested for an association using a recessive model where a one-sided p-value < 0.05 was considered statistically significant.CEP72 rs924607 TT homozygotes .We were unable to confirm the association between the rs924607 TT genotype and neurotoxicity  in our global meta-analysis. Analysis of the continuation phase (following induction) studies showed significantly higher odds for neuropathy in  centrosomal protein 72 is associated with more neurological toxicity in children treated with vincristine for acute lymphoblastic leukemia. We found that this variant was indeed connected with higher odds for developing peripheral neuropathy in the later stages of therapy.We analyzed the findings of four previous studies to find out if a variant of a gene important for in cell division called  Despite great advances achieved in the last few decades, cancer remains the leading cause of child death by disease in developed countries. Acute lymphoblastic leukemia (ALL) is the predominant childhood malignancy, representing more than 30% of all childhood cancers and approximately 80% of all childhood leukemias. Fortunately, its cure rates have surpassed 85% ,2. As moThe efficacy of current ALL treatment has been achieved mainly by optimizing the use of existing combinations of chemotherapeutics, rather than adding novel drugs . TherefoVincristine (VCR) is a very widely used anticancer drug, utilized in pediatric treatment protocols for leukemia as well as solid tumors . It is aCEP72); an overview of other potential biomarkers is given in Several studies have assessed various candidate genes involved in VCR metabolism; however, they have not uncovered genetic variants consistently linked with a higher risk of VIPN . The majCEP72 gene. A genome-wide association study (GWAS) by Diouf and colleagues established the rs924607 TT as a VIPN risk genotype [CEP72 encodes a protein crucial for centrosome formation. Using expression studies, Diouf et al. showed that TT homozygotes had significantly lower CEP72 mRNA levels. They were able to explain this effect by demonstrating that the rs924607 T variant creates a binding site for the NKX-6.3 transcription repressor in the promoter region of the gene. Furthermore, they created pre-B ALL, T-ALL and neuroblastoma cell models and employed short hairpin RNA to impair CEP72 gene expression. These cells were then shown to be more sensitive to VCR [A promising VIPN biomarker is a genetic variant in the promoter of the et al. and Zgheib et al. reporting no association between CEP72 rs924607 and VIPN and Wright et al. reporting a significantly higher risk of VIPN linked with TT genotype  [et al. limited their study of VIPN to remission induction phase, while the other three studied VIPN across all phases of treatment. Some authors have also highlighted the importance of precise definitions of study end points. Namely, VIPN is most commonly diagnosed and graded using the National Cancer Institute Common Terminology Criteria for Adverse Events which classifies grade 1 as mild, grade 2 as moderate, grade 3 as serious/disabling and grade 4 as life-threatening neurotoxicity. However, these clinical scales are dependent on self-reporting of symptoms which is particularly problematic in children [This study was followed by four replication studies, three of which were conducted on pediatric ALL patients and one on adult ALL patients ,23. The 3–12.66) . Such inchildren . NeverthCEP72 gene and VIPN [A recent GWAS study found no association between the and VIPN . This stCEP72 TT genotype (estimated to comprise 16% of the study population) and shorter duration of therapy in patients with CEP72 CC or CT genotype will decrease the incidence and/or severity of VIPN. The researchers are using an unblinded design, where the treating physicians will be aware of the treatment assignments, but the investigators who evaluate neuropathy will not. The study is currently still recruiting patients [There is also a promising ongoing clinical trial called Total Therapy XVII for Newly Diagnosed Patients With Acute Lymphoblastic Leukemia and Lymphoma sponsored by St Jude’s Research Hospital  whose goal is to use strategies based on specific genomic features to improve outcomes for children with ALL and acute lymphoblastic lymphoma . They arpatients .CEP72 rs924607 and VCR toxicity we decided to conduct a meta-analysis of the results of studies on pediatric patients treated for ALL.In an attempt to elucidate the association between CEP72 rs924607 genotype TT, we searched the PubMed database for research articles by using the following keywords: ‘CEP72 rs924607 and Vincristine’, ‘CEP72 rs924607 and neuropathy’, ‘CEP72 rs924607 and Vincristine andneuropathy’ as well as ‘CEP72 rs924607 and neurotoxicity’ and ‘CEP72 rs924607 and pediatric cancer’. Subsequently, we searched for cross-references.To conduct the meta-analysis of data on the association between VIPN and CEP72 rs924607 genotyping data. We excluded studies that were done on adult patients and those that were not written in the English language. We did not exclude studies based on the publication date.Our inclusion criteria were that the study had to report the incidence of VIPN in a pediatric population treated for ALL and also include the From a total of six articles retrieved on PubMed, two were excluded. One was a review article that did not include genotyping data and the other a study on adult patients ,23. The CEP72 rs924607 genotype, incidence of VIPN, VIPN grade, duration of therapy before VIPN occurred and cumulative VCR dose received before VIPN.From the four selected articles we extracted data on the Symptoms of VIPN were defined as neuropathic symptoms and sensory, motor and autonomic dysfunction . Peripheral neuropathy was graded according to the NCI Common Terminology Criteria for Adverse Events version 1.0. Patients with grades 2, 3 or 4 were considered VIPN cases. Occurrences of VIPN were the end points used for this analysis.CEP72 rs924607 VIPN risk genotype TT, we tested for an association between this marker and VIPN using a recessive model. A one-sided p-value < 0.05 was considered statistically significant. We performed a random-effects meta-analysis using the RevMan software [CEP72 rs924607 TT genotype and VIPN was separately estimated in a subgroup of studies that assessed VIPN incidence in the continuation phases of ALL therapy. The continuation phase is defined as a treatment phase following remission induction therapy.To investigate the software . The OR This study encompasses 315 cases of VIPN from four studies . AccordiCEP72 rs924607 TT homozygotes was 1.99   .et al. [2 = 58%, p = 0.09).When considering only continuation phases studies , the ORet al. established CEP72 rs924607 as a promising VIPN biomarker, by reporting a statistically significant association between CEP72 rs924607 TT risk genotype. We analyzed the data from four such studies and revealed the association between the CEP72 rs924607 TT genotype and VIPN in our continuation phases analysis. However, our global analysis did not confirm such association  studied the North American children while the study by Zgheib et al. was done in the Saudi population and the study by Gutierrez-Camino et al. on the Spanish population [It is worth noting that the studies were conducted on populations with distinct genetic backgrounds. Both studies that confirmed the association  clinical trial which is a prospective study and will hopefully further clarify the role of practice . We hopeCEP72 in VCR metabolism is obtaining confirmatory evidence through functional studies. In the discovery study, Diouf and colleagues attempted to do this by impairing the expression of the gene in pre-B ALL, T-ALL and neuroblastoma cell lines and then subjecting those cells to VCR. They found that these cells were more sensitive to VCR. Furthermore, the CEP72 promoter single nucleotide polymorphism rs924607 genotype T was shown to create a binding site for a transcriptional repressor thus reducing CEP72 expression in human neurons and leukemia cells and increasing their sensitivity to VCR [Another important future research direction for the role of CEP72 rs924607 genotype TT is likely associated with neurotoxicity occurring during the continuation phases (after induction) of pediatric ALL treatment. However, further studies will be needed before we can implement this finding at the bedside and improve patient outcomes in the future.In the present study, we managed to show that As high-throughput technologies are providing us with increasingly high-quality data about genomic and transcriptomic attributes of ALL patients and their response to therapy, we are bound to discover even more VIPN biomarkers. In 5–10 years, we hope to have a reliable biomarker for VIPN for the clinical setting, which will allow us to attenuate therapy with VCR in at-risk patients. This will improve both their short-term and long-term quality of life.CEP72 rs924607 has great potential to become such a biomarker. The Total Therapy XVII for Newly Diagnosed Patients With Acute Lymphoblastic Leukemia and Lymphoma (NCT03117751) clinical trial should provide us with additional answers on this topic. However, in the future, we hope to see more functional studies that will help us understand the mechanisms that cause VIPN and eventually prevent it.CEP72 rs924607 TT genotype and neurotoxicity in our global meta-analysis.We were unable to confirm the association between the CEP72 rs924607 TT homozygotes.The analysis of the continuation phase studies exposed a significantly higher risk for neuropathy in CEP72 rs924607 TT genotype and vincristine-induced peripheral neuropathy exclusively in pediatric patients.Our meta-analysis is the first validation of the association between the Because we replicated this finding in such a diverse pediatric acute lymphoblastic leukemia cohort, this pharmacogenomic biomarker could be a valuable tool for clinicians.Developing such biomarkers is a path toward personalized medicine.


Three associations were graded as moderate, and seven associations were rated as weak. This study presents the relationships between SNPs in miRNAs or biosynthesis genes and LC risk, subsequently demonstrates the credibility of these significant associations, and highlights the role in the pathogenesis of LC.In the past decade, the studies involving single nucleotide polymorphisms (SNPs) in microRNAs (miRNAs) with lung cancer (LC) risk have been performed, however, these results are inconsistent, and a systematic research synopsis has not been performed yet. Therefore, we attempted to perform comprehensive meta‐analyses to assess the relationships between SNPs in miRNAs or biosynthesis genes and LC risk and further evaluate the epidemiological credibility of these significant associations. We used PubMed, Medline, and Web of Science to search for relevant articles published before 30 May 2019 that assessed relationships between SNPs in miRNAs or biosynthesis genes and LC risk. The cumulative epidemiological evidence of statistical relationships was further assessed combining Venice Criteria and a false‐positive report probability test. Based on 20 publications with 15 969 cases and 17 174 controls, we found that six variants in miRNAs or biosynthesis genes that proved significant associations with LC risk, whereas five proved no association. Subgroup analyses by ethnicity and genetic models were performed, suggesting that four associations were rated as demonstrating strong evidence of relationship with LC risk, including  Our study evaluated the cumulative evidence of significant associations between SNPs in miRNA or biosynthesis genes and lung cancer (LC) risk by combining Venice Criteria and false‐positive report probability test. Two variants with four associations were rated as demonstrating strong evidence with risk of LC, three associations were moderate, and seven associations were weak. Our paper provides summary evidence with relationships between SNPs in miRNAs or biosynthesis genes and LC risk, subsequently demonstrates the credibility of these significant associations, and highlights the role in the pathogenesis of LC. In addition, the references in included articles were also checked to obtain other potential relevant data.2.2The following criteria should be considered with selection of studies: (a) investigating relationships between miRNA‐SNPs or biosynthesis genes and risk of LC with studies performed in a case‐control or cohort design in human; (b) patients with LC were pathologically or histologically confirmed; (c) presenting the sample size in cases and controls; wherever necessary, the amount of genotype and/or allelic distributions should be offered; (d) the published or online full text in the journals was in English. Studies were excluded if: (a) studies had insufficient relevant data; (b) studies were published by form of conference abstracts and letters to editors, rather than as full reports; (c) the studies were mainly based on LC survival/mortality rate (rather than incidence).2.3https://www.ncbi.nlm.nih.gov/snp/) and then used the most recent one. We used the Newcastle Ottawa Scale (NOS) to evaluate the quality of included eligible studies.The relevant information was independently extracted by two authors (GL and JT) and were crosschecked each other. Any disagreement was discussed with the third investigator (HC) and finally resolved together. For the qualified SNPs, the following publication details were extracted, including first author, the year of publishing, study design, ethnicity, gene name, variation in gene, the sample size in cases and controls, minor allelic frequency (MAF), genotype counts for cases and controls. Specifically, two major ethnicities, Asian and Caucasian, were frequently reported in our study; “all populations” indicate two or more. If the same study population was reported in more than one article, we collected data from the most recently published study with the greatest number of and most integrated participants. As for the same genetic variant, the modes of presentation were inconsistent, we therefore first checked on the website . Additionally, we adopted the random effect model if the P‐value was <.1, otherwise the fixed effect model was used. Moreover, we performed sensitivity analyses for all SNPs, especially for SNPs with significant associations, further to assess whether the significant ORs were robust by excluding a single study (dataset), or the first published study, or studies deviated from the Hardy‐Weinberg equilibrium (HWE) in the controls. Begg's test and Egger's test were used to evaluate potential publication bias and small study bias, respectively.P‐value <.05 in the meta‐analysis and <.1 in the Cochran's Q, Begg's and Egger's tests, as well as a test for excess significant findings, was considered significant, as recommended.We carried out comprehensive meta‐analyses using allelic, dominant, and recessive models (See Table 2.5I2 ≤ 25%), grade B (25% < I2 < 50%), or grade C (I2 ≥ 50%). Protection from bias was mainly determined using sensitivity analysis and a series of bias tests including publication bias and small‐study bias, as well as an excess of significant findings. Generally, grade A was assigned to indicate no observable bias, or if bias was unlikely to explain the presence of the association, grade B was assigned if bias could be present, and grade C was allocated if the bias was evident or was likely to explain the presence of the association. We utilized an extensive checklist to check the sources of bias in different settings proposed by the Venice Criteria (see Supporting Information notes). The magnitude of the association was related to the evaluation of protection from bias; a summary OR < 1.15 (or >0.87 in a protection effect) was graded as a score of C for an association, unless the association had been replicated, prospectively, by several studies with no evidence of publication bias .First, the Venice Criteria were used to evaluate the epidemiological credibility of significant associations identified by the meta‐analyses.As Wacholder et al suggested, a prior probability of 0.05 and a false‐positive report probability (FPRP) cutoff value of 0.2 in FPRP assay should be performed to detect the potential false‐positive results among significant associations and assess whether these associations should be omitted.33.1Our search yielded a total of 596 publications. As presented Figure 3.2miRNA‐146a rs2910164, miRNA‐499 rs3746444, and miRNA‐27a rs895819) and three SNPs on miRNA‐related biosynthesis genes . Specifically, our study demonstrated a significant association between miRNA‐146a rs2910164 and LC risk in all populations , and in Asians , but demonstrated no significant association in Caucasians under the three genetic models. For miRNA‐499 rs3746444, we found that SNP rs3746444 under dominant model had significant association with LC risk in all populations , and a nominally significant association between SNP rs3746444 and risk of LC was found in Asians under recessive model , rather than the allelic or the dominant model, as well as in Caucasians under all three models.We conducted meta‐analyses to assess the relationships between the miRNA‐SNPs or biosynthesis genes and LC risk. These results are presented in Table AGO1 rs595961 under the allelic and the recessive models , rather than the dominant model. For miRNA‐27a rs895819, our study presented that SNP rs895819 had a nominally significant association with LC risk under the recessive model , but not the allelic or the dominant model. For GEMIN4, two SNPs (rs7813 and rs910924) had significant association with LC risk; the former SNP had significant association with an increased risk of LC under the allelic model , rather than the dominant or the recessive model; the latter variant had significant association with an decreased risk of LC under the allelic and the dominant models , rather than the recessive model.The rest of four SNPs significantly associated with LC risk were exclusively performed in Asians; significant associations were observed in miRNA‐196a2 rs11614913, miRNA‐4293 rs12220909, miRNA‐149 rs2292832, miRNA‐608 rs4919510, and miRNA‐26‐1 rs7372209; of these five variants, two variants were performed both in Asians and Caucasians and three variants concentrated on single (Asian) population.In addition, current study found that five variants in five miRNAs had no significant association with LC risk under all three genetic models, including 3.3miRNA‐146a rs2910164 in all populations under dominant model and in Asians under the dominant and the recessive models; AGO1 rs595961 in Asians under the allelic model), moderate for three associations , and weak for seven associations  based on Venice Criteria.Cumulative epidemiological evidence was graded for six SNPs associated with LC risk; the details of the evidence are presented in Table P‐value of FPRP assay less than .05 for two associations , from .05 to .2 for nine associations, and greater than .2 for the rest of three associations . Therefore, cumulative epidemiological evidence of an association was rated as strong for miRNA‐146a rs2910164 in all populations under the dominant model and in Asians under the dominant and the recessive models, and for AGO1 rs595961 in Asians under the allelic model  in all populations, rs2910164  in Asians, rs595961  in Asians, and rs910924  in Asians; moderate heterogeneity was detected for relationships of rs2910164  in all populations, rs2910164  in Asians, rs3746444  in Asians, and rs895819  in Asians; large heterogeneity was found for relationship of rs3746444  in all population, and rs7813  in Asians. There was little evidence of publication bias for associations of SNPs with LC risk , except for rs2910164 in all populations under the allelic and the recessive models, and in Asians under the allelic model (P < .10). In addition, we performed sensitivity analyses to assess the robustness of results of these significant relationships and found that removal of a single study (dataset), the first published or studies deviated from HWE in controls did not change the summary ORs (data not shown), except for the rs895819 in Asians under the recessive model and the rs3746444 in all populations under the dominant model and in Asians under the recessive model. In our sensitivity analyses, after the exclusion of study deviated from HWE in controls, no significant association was observed in the three genetic models above. In addition, three significant associations did not remove a single study (dataset) due to only two primary datasets, including rs595961, rs7813, rs910924 in Asians under allelic model and rs910924 in Asians under the dominant model  were upgraded from weak to moderate (FPRP < 0.05); a publication bias may explain how the former associations were graded “ABC” overall based on Venice Criteria; these associations were thus rated as having moderate associations with LC risk because of a FPRP value <0.05. The latter variant (SNP rs59561) was not upgraded or downgraded based on FPRP value (0.05 < FPRP < 0.2); the criteria with amount of evidence may explain how this association was graded “BAA” overall based on Venice Criteria. However, this variant was conducted exclusively in Asians. One reason that concentrated on single (Asian) population could be the small sample size of this meta‐analysis, which made subgroup analyses challenging. Therefore, further expanding sample size and assessment on other ethnicities into this variant are necessary.Three associations were graded as demonstrating moderate evidence of relationship with risk of LC, including miRNA‐146a rs2910164 under the recessive model and miRNA‐499 rs3746444 under the dominant model were considered as being significant association with LC risk in all populations, while other five associations were significantly associated with LC risk in Asians, including miRNA‐499 rs3746444 under the recessive model, GEMIN4 rs7813 under the allelic model, miRNA‐27a rs895819 under the recessive model and GEMIN4 rs910924 under the allelic and the dominant models. Of these variants, GEMIN4 rs7813 increasing LC risk by 1.258‐fold in the meta‐analysis was well established, with an overall schema of ACA, indicating a high degree of heterogeneity that may explain how this variant was rated as weak based on Venice Criteria; no statistical data could be extracted for ethnicity subgroups, which could explain the heterogeneity in the data. We recommend subdividing populations by ethnicity to identify potential differences in the association between this variant and LC risk. Our study presented that miRNA‐27a rs895819 had no association with LC risk in all populations. Interestingly, we found that SNP rs895819 was significantly associated with LC risk in Asians , rather than in Caucasians . While sample size may be one factor affecting different associations between ethnicities, other factors such as methodology, LC subtypes, and environmental factors may also as account for variation in the data. Moreover, two variants, miRNA‐146a rs2910164 in all populations under the recessive model and miRNA‐27a rs895819 in Asians under the recessive model, were rated as weak, with an overall schema of ABC and BBC, respectively; the publication bias was main factor triggering the weak evidence for this variant, further investigations on these studies were recommended. Additionally, GEMIN4 rs910924 under the allelic and the dominant models in Asians was rated as weakly associated with LC risk , after being downgraded from moderate to weak based on FPRP value (>0.2). Previous studies showed that setting different prior probabilities may make their results more noteworthy. Results therefore would be more convincing if different prior probabilities were assigned based on FPRP assay. Moreover, it may be necessary to further expand the sample size on current study and on other ethnicities into these variants, and then investigate these associations in greater depth.Seven associations were rated as being weakly associated with LC risk. Among these seven associations, miRNA‐146a rs2910164 and miRNA‐499 rs3746444 with 101 cases and 129 controls, respectively; the small sample size for non‐Asian race may make subgroup analyses challenging. For genetic models, three models were used for more comprehensive evaluation, while the existence of different genetic background such as age and gender about patients, subtypes of LC and environmental factors such as cigarette smoking were not be taken into consideration may present as sources of variation in the result. Further study on these factors is recommended.Additionally, the associations stratified by different ethnicity or genetic models were usually inconsistent. For ethnicity, all studies were performed on a single ethnic group (Asian) except miRNA‐4293 rs12220909, miRNA‐149 rs2292832, miRNA‐608 rs4919510 and miRNA‐26‐1 rs7372209 and miRNA‐196a2 rs11614913; the former four SNPs were observed in a sample of approximately 4000 cases, at approximately 95% power to detect an OR of 1.15 in an allelic model for a variant with MAF of 20%. In addition, the last variant was observed in a sample of approximately 10 000 cases, at approximately 98% power to detect an OR of 1.15 in an allelic model for a variant with MAF of 20%. Therefore, these five variants may be not associated with LC risk. It is probable that further investigations evaluating these five variants will not yield fruitful results with LC risk if the sample size less than current study.Our study presented that five variants had no association with LC risk under all three genetic models, including Some limitations should be considered to this study: (a) although available studies were searched widely, some publications may have been missed; (b) the excess of significant findings was not further evaluate due to insufficient data; (c) subgroup analyses were exclusively performed by ethnicity (only for Asians and Caucasians) and genetic models , which may make the credibility of some results challenging, especially for weak evidence with small sample size; future evaluation with much larger sample size and other ethnicities may be necessary to confirm or refute these connections; (d) we only assessed the susceptibility/incidences of associations between SNPs in miRNA or biosynthesis genes and LC risk; the involvement of genetic polymorphisms as they contribute to cancer progression, metastasis, and drug resistance in LC was not evaluated due to insufficient data. Despite these limitations, we believe that this paper, which provides a comprehensive summary and evaluation of existing literature on the role of SNPs in miRNA or biosynthesis genes to LC, will be of value in informing future genetic studies.In our study, Venice Criteria and FPRP test were introduced to evaluate the cumulative evidence of significant associations to increase the persuasion and accuracy of the final results. Two variants with four associations were rated as demonstrating strong evidence with risk of LC, three associations were moderate, and seven associations were weak. The results of SNPs in miRNAs or biosynthesis genes with LC risk may help us to get more potential target population for primary prevention. In summary, our study summarizes current literature on the SNPs in miRNA or biosynthesis genes architecture of LC susceptibility, and provides useful information for designing future studies aiming to evaluate SNPs in miRNA or biosynthesis genes factors for LC risk.The authors have no conflict of interest. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file. Click here for additional data file.


This study aimed to quantitatively summarize the evidence for VDR BsmI gene polymorphism and osteoporosis risk in postmenopausal women.The PubMed, EMBASE, Weipu, CNKI, and Wanfang databases were searched for eligible studies. Case-control studies containing available genotype frequencies of B/b were chosen, and odds ratio (OR) with 95% confidence interval (CI) was used to assess the strength of this association.p = 0.019; recessive model: OR = 0.736, 95% CI 0.568~0.955, p = 0.021; and co-dominant model: bb vs. BB OR = 0.701, 95% CI 0.511~0.962 p = 0.028), and we failed to find any significant relationship in Asians.4485 osteoporosis and 5490 controls were identified in our meta-analysis. In the stratified analysis, a significant association was observed between VDR BsmI gene polymorphism and osteoporosis susceptibility in Caucasians (additive model: OR = 0.809, 95% CI 0.678~0.965, The present meta-analysis suggests that VDR BsmI genotype is associated with increased risk of postmenopausal osteoporosis in Caucasians but not in Asians. To draw comprehensive and true conclusions, further prospective studies with larger numbers of participants worldwide are needed to examine associations between VDR BsmI polymorphism and osteoporosis in postmenopausal women. Osteoporosis, as a systemic bone disease characterized by decreased bone mineral density, micro-structure deterioration of bone tissue, and increased risk of bone fracture , 2, is cThe interaction between vitamin D and its receptor exerts an important role in calcium homeostasis and bone metabolism by regulating osteocyte growth and differentiation, intestinal calcium absorption, and parathyroid hormone secretion . The vitPostmenopausal osteoporosis, resulting from estrogen deficiency, is the most common type of osteoporosis, and estrogen deficiency results in an increase in bone turnover owing to effects on all types of bone cells . In 1996http://www.ncbi.nlm.nih.gov/pubmed), EMBASE (http://www.embase.com), Weipu (http://www.cqvip.com/), CNKI (http://www.cnki.net/), and Wanfang (http://g.wanfangdata.com.cn/) databases were thoroughly searched by the authors . The keywords were “vitamin D receptor” or “VDR” and “osteoporosis” or “fracture” and “BsmI” or “rs1544410” in combination with “genetic” or “polymorphism” or “variant”.PubMed  of lumbar spine and femoral neck, osteoporosis was defined as BMD ≤ − 2.5 SDs (T-score).Exclusion criteria of this meta-analysis are listed as follows: ① reviews, case reports, comments, and letters; ② incomplete data; ③ without full text. In addition, all relevant references were also reviewed. If there were duplicate data in papers published by the same author, only the most recent or complete study was included in this analysis.Two independent investigators extracted data from eligible studies; the characteristics included the following: ① the 1st author, ② publication year, ③ region, ④ ethnicity, ⑤ age range, ⑥ sample size, ⑦ allele frequency of cases and controls, and ⑧ genotyping method. Any different evaluation results need to be revisited until a consensus is reached.The quality of eligible publications was assessed by the Newcastle-Ottawa quality assessment scales (NOS) . The scaΧ2 test. The gene frequencies of the control group must conform to the Hardy-Weinberg equilibrium (p > 0.05). The relationship between VDR BsmI gene and osteoporosis was accessed by calculating odds ratios (ORs) and 95% confidence intervals (CIs). The pooled ORs were performed for additive genetic model (b vs. B), dominant model (bb + Bb vs. BB), recessive model (bb vs. Bb + BB), and co-dominant model  respectively. The subgroup analyses by ethnic groups were also performed. The statistically significant p value was set at 0.05. Heterogeneity assumption was evaluated by a chi-square-based Q test (p < 0.05 indicated heterogeneity across studies). The summary OR estimate of each study was calculated by the fixed-effects model if there was no significant heterogeneity. Otherwise, the random-effects model was used [p < 0.05 considered representative of statistical significance) [The observed genotype frequencies of the VDR BsmI polymorphism in control groups were assessed for Hardy-Weinberg equilibrium using the was used , 17. Theficance) . All anaLiterature screening process is shown in Fig. p = 0.019; recessive model: OR 0.736, 95% CI 0.568~0.955, p = 0.021; and co-dominant model: bb vs. BB OR 0.701, 95% CI 0.511~0.962, p = 0.028, Fig. Differences in allelic distribution by ethnicity could be partially responsible for the observed differences in the association between VDR BsmI and osteoporosis. The evaluations of the association between VDR Bsml polymorphism and osteoporosis risk in postmenopausal women are summarized in Table p = 0.856; b/b vs. Bb/BB: p = 0.851; Bb/bb vs. BB: p = 0.813; Bb vs. BB: p = 0.510; bb vs. BB p = 0.937).Begg’s test was performed to quantitatively evaluate the publication bias of literatures on osteoporosis. The results provided statistical evidence in overall results, suggesting the absence of publication bias. All graphical funnel plots of the included studies appeared to be symmetrical. There was no visual evidence of publication bias visually from the funnel plot, which implied that the publication bias was low in the present overall meta-analysis (b vs. BB: Genetic difference is one important factor affecting the susceptibility to osteoporosis. VDR gene has been widely studied because of its important role in regulating bone metabolism and bone homeostasis. The VDR Bsml polymorphism is located in the 3′ untranslated region (UTR). It is involved in regulating the stability of VDR mRNA and is one of the most important subtypes of VDR gene polymorphism. Studies on VDR Bsml polymorphism and susceptibility to osteoporosis are various, but the results are not consistent. A recent meta-analysis  shows thIn conclusion, our study believes that VDR BsmI polymorphism and postmenopausal osteoporosis are genetically linked in Caucasians, but not in Asians. It is necessary to conduct large-scale studies to verify the correlation of different populations and environmental factors in the susceptibility to osteoporosis.


Background: Platinum-based agents, including cisplatin, carboplatin, and oxaliplatin, are indispensable for the treatment of lung cancer. The development of toxicity frequently necessitates dose reduction or discontinuation of therapy, despite the clinical response. Pharmacogenomics studies were reviewed to identify the possible genetic variants that underlie individual susceptibility to platinum-related toxicities.Method: We conducted a systematic search in PubMed and Embase for pharmacogenomics reports that focused on commonly reported platinum-induced toxicities, such as gastrointestinal (GI), hematological, neurological, and other toxicities, in patients diagnosed with lung cancer. Meta-analyses were conducted to determine the association between genetic polymorphisms and platinum-induced toxicity by checking the odds ratio (OR) and 95% confidence interval (CI) using random or fixed-effects models as appropriate.Results: Twenty eligible studies that met the inclusion criteria with sufficient data were extracted and presented comprehensively. A total of 16 polymorphisms from 11 genes were included in the meta-analysis. MTHFR rs1801131 and MDM2 rs1690924 were significantly correlated with platinum-induced GI toxicity . Patients with the MTHFR rs1801131AA and MDM2 rs1690924TC/CC genotype tended to have a higher risk of GI toxicity than patients with other genotypes did . Compared to carriers of the MTHFR rs1801133CC genotype, carriers of the CT/TT genotype had a significantly increased risk of hematological toxicity .Conclusion: In the future, physicians should pay careful attention to MTHFR and MDM2 for personalized chemotherapy treatment among patients with lung cancer. Lung cancer is the second most commonly diagnosed malignant tumor in men and women and is one of the main causes of mortality worldwide . There aPlatinum-based agents, including cisplatin, carboplatin, and oxaliplatin, in combination with other cytotoxic drugs have been recommended as the first-line chemotherapy for lung cancer . The antGenetic polymorphisms contribute to the differences in platinum-related toxicities, and there is accumulating evidence to support this speculation . TherefoERCC1 C118T was significantly associated with platinum-induced toxicity while other studies presented contradictory results  to summarize the pharmacogenomics of platinum-based toxicities in patients with lung cancer and (2) to provide a comprehensive assessment of the association between genetic polymorphisms and platinum-based drug response in patients with lung cancer. We collected all available publications on pharmacogenomics studies that platinum-based toxicities in patients with NSCLC and SCLC and quantitatively studied them using a meta-analysis strategy.For the literature search, two authors (Z. Y. Luo and W. H. Liu) independently performed a systematic literature search in three databases: PubMed database, Cochrane Library, and ISI Web of Knowledge. The search results were reviewed and compared by a third reviewer (Y. Zheng), and discrepancies between searchers were discussed and solved with consensus. The literature was searched from the first available article to June 18, 2019. Publications were retrieved using terms associated with platinum drugs  in combination with keywords associated with genetic variation (“polymorphism” or “SNP” or “single nucleotide polymorphism” or “mutation” or “variation”) and “toxicity” or “adverse effect,” and “lung cancer.”All identified abstracts were carefully and independently reviewed by two investigators (Z. Y. Luo and W. H. Liu) for eligibility. The inclusion criteria were as follows: (i) clinical studies, regardless of the sample size; (ii) studies that assessed the associations between genetic polymorphisms and platinum-induced severe (grade 3–4) toxicities in patients with lung cancer, and toxicities were evaluated and graded according to the Common Terminology Criteria for Adverse Events; (iii) numbers for each genotype were available or could be calculated in different groups; and (iv) at least two studies that evaluated one polymorphism met the abovementioned three criteria. If the two investigators disagreed about the eligibility of an article, it was resolved by consensus with a third reviewer (J. Q. Luo).Data were manually extracted by two reviewers  who were blinded to each other and used the same data recording form. All data were reviewed by the third reviewer (J. Q. Luo) until they reached a consensus on all of the data extraction items. The following information was extracted from each study: name of the first author, publication year, country of the study and ethnicity of the patients, sample size, tumor type, disease stage, chemotherapeutic drugs, platinum-induced toxicities, toxicity evaluation criteria, genes and Single Nucleotide Polymorphism Database number of the investigated polymorphisms, and genotype methods.Z-test. The Cochran's Q-test and I2 metric were performed to determine the possibility of between-study heterogeneity. The heterogeneity of publications in each meta-analysis was considered to be significant at P < 0.05 for the Q statistics and I2 > 50% for the I2 metric. Sensitivity analysis was conducted if publication bias existed; one study was excluded at a time, and the others were analyzed to estimate whether the results were affected markedly by a single study. Subgroup analysis based on toxicity  was conducted to assess the sources of heterogeneity across the studies. Potential publication bias was assessed using the Begg and Egger tests, with P < 0.05 considered to indicate a significant publication bias. All statistical analyses were performed with by the Cochrane Collaboration software .The patients were divided into two groups: those with grade 3–4 (severe) toxicities and those with grade 0–2 (no or mild) toxicities. The pooled odds ratio (OR) and associated 95% CI were calculated and used to evaluate the strength of association in different genotype groups. The significance of the pooled estimates of the OR was determined using the The initial research yielded 1,003 publications. A total of 608 articles were excluded owing to duplicate publication; in addition, 306 studies were excluded in the first round of review, among which 30 were irrelevant literature articles, 5 were not English articles, 64 were reviews or meta-analysis, 27 were case reports or abstracts, 19 were non-clinical-related studies, 13 focused on other tumors, 41 did not evaluate platinum-based chemotherapy, 20 did not focus on platinum-related toxicities, and 89 were not pharmacogenomic studies. After reading the full text of these articles, we found that 12 studies did not provide detailed data owing to the lack of significant association between gene polymorphisms and platinum-based toxicities.Finally, 20 publications with sufficient data met the inclusion criteria and were extracted . The genThe main results of this meta-analysis showing the association between polymorphisms and risk of platinum-based grade 3–4 toxicities are shown in ERCC1 C118T. The association between C118T polymorphism and platinum-based grade 3–4 hematological and GI toxicity was found and replicated by six and three studies, respectively. We first performed a meta-analysis to determine the association between C118T polymorphism and grade 3–4 hematological toxicity by including 1,450 subjects. No significant relationship was detected between C118T polymorphism and grade 3–4 hematological toxicity  using a fixed-effect model. On subgroup analyses based on ethnicity, the combined OR for risk and the I2 were consistent and did not show any apparent fluctuation (The most extensively studied polymorphism was ctuation .P = 0.38) using a fixed-effect model. This association did not significantly change on subgroup analyses based on ethnicity.We further analyzed the relationship between C118T polymorphism and grade 3–4 GI toxicity by including 790 patients. Pooled data from these investigations showed grade 3–4 GI toxicity rates of 13.49 and 14.15% in the CC + CT genotype and the TT genotype groups, respectively. No significant relationship was detected between C118T polymorphism and grade 3–4 GI toxicity  using a fixed-effect model. This association had no significant change on subgroup analyses based on ethnicity. Among 196 patients who were included to determine the association between C8092A polymorphism and grade 3–4 nephrotoxicity, no significant association was observed  using a fixed-effect model.A total of four and two studies examined the association between C8092A mutation and platinum-based grade 3–4 hematological toxicity and nephrotoxicity, respectively. No significant association was detected between C8092A polymorphism and grade 3–4 hematological toxicity  using a random-effect model (I2 = 71%). After removing the study by Peng et al., the publication bias disappeared (I2 = 10%). We further examined the raw data of this study and found that the G1196A mutation was not considered in the Hardy–Weinberg equation (P = 0.002). No significant association was detected between G1196A polymorphism and grade 3–4 hematological toxicity  using a fixed-effect model. On subgroup analyses based on ethnicity, the combined risk of G1196A mutation on platinum-based grade 3–4 GI toxicities was consistent and did not show any apparent fluctuation  using a fixed-effect model. This association did not significantly change on subgroup analyses based on ethnicity  using a random-effect model. In addition, publication bias disappeared (I2 = 0%) after the study by Wang et al. was removed. The possible reason is that the call rate for G2677T/A genotypes was <90% in their study, which could have led to inaccuracy in their results. On subgroup analyses based on ethnicity, different ethnic populations showed distinct effects for the G2677T/A polymorphism. There were significant protective effects of the T/A allele on the risk of platinum-based overall toxicity in the non-Chinese subgroup  using a random-effect model. After removing the study by Qian et al., publication bias disappeared (I2 = 17%). Furthermore, the association between the G2677T/A polymorphism and grade 3–4 GI toxicity was analyzed. No significant association was detected between the G2677T/A polymorphism and grade 3–4 GI toxicity  using a random-effect model. In addition, the publication bias vanished (I2 = 0%) when the study by Qian et al. was removed. The study by Qian et al. was a high-quality well-designed study, with the appropriate sample size. The association between the G2677T/A polymorphism and grade 3–4 hematological and GI toxicities was still non-significant after the study by Qian et al. was removed.Moreover, no significant association was detected between the G2677T/A polymorphism and grade 3–4 hematological toxicity  using a random-effect model. In addition, the publication bias vanished (I2 = 0%) when the study by Wang et al. was removed. The call rate for the −24C>T genotype was 83% in this study, which could have led to inaccuracy in the result. On subgroup analyses based on ethnicity, the association between the −24C>T polymorphism and grade 3–4 hematological toxicity was still non-significant  using a fixed-effect model. On subgroup analyses based on ethnicity, the association between the A313G polymorphism and grade 3–4 hematological toxicity was still non-significant  using a fixed-effect model. Three studies examined the association between the G934A polymorphism and grade 3–4 hematological toxicity. No significant association was detected between the G934A polymorphism and grade 3–4 hematological toxicity  using a random-effect model. The publication bias disappeared (I2 = 0%) when the study by Tibaldi et al. or that by Powrozek et al. was removed. The main reason for this phenomenon was the small sample size of these two studies.A total of four studies determined the association between the A2251C polymorphism and grade 3–4 hematological toxicity. No significant correlation was detected between the A2251C polymorphism and grade 3–4 hematological toxicity  on a fixed-effect model. However, no significant association was found between the A1298C polymorphism and grade 3–4 hematological toxicities  using a fixed-effect model.In total, three included studies evaluated the association between the A1298C polymorphism and grade 3–4 GI and hematological toxicities. Carriers of the AA genotype had more severe GI toxicity than carriers of the AC + CC genotype did  using a fixed-effect model. The pooled data showed that patients with the 677CC genotype had an increased risk of severe hematological toxicity than the carriers of the CT + TT genotype did  using a fixed-effect model. The association between these two polymorphisms and grade 3–4 hematological toxicity was unchanged on subgroup analyses based on ethnicity  using a random-effect model. Similarly, we investigated the influence of a single study on the overall risk by excluding one study at a time. However, the combined overall risk and I2 were consistent and did not show any apparent fluctuation.Four studies with 1,189 Chinese patients evaluated the correlation between the rs2279744 polymorphism and grade 3–4 hematologic toxicity. There was no significant correlation between the rs2279744 polymorphism and grade 3–4 hematological toxicity  using a fixed-effect model.Two studies were used for analyzing the association between the rs1690924 polymorphism and platinum-based toxicities, including grade 3–4 overall, hematological, and GI toxicities. The pooled results showed no significant correlations between the rs1470383 polymorphism and grade 3–4 overall and hematological toxicities. The pooled results from all patients indicated that the carriers of the TT + TC genotype had a markedly increased risk of grade 3–4 GI toxicity than carriers of the CC genotype did  is a proto-oncogene and a key negative regulator of p53. MDM2 plays a role in P53-independent antitumor activity through directing binding, ubiquitination, and degradation of the p53 gene (MDM2 gene polymorphism (rs2279744) was associated with the risk of lung cancer and the clinical outcomes  pathway are among the most clinically relevant genetic determinants with susceptibility to platinum-based toxicities . The NER and XPD . In factAlthough previous studies validated that polymorphisms in genes act as potential risk factors for platinum-induced toxicities considering individual differences, the results of our research indicated that some polymorphisms correlated with platinum-induced toxicities had limited contribution to the interindividual differences in platinum-induced toxicities. The reasons for negative or conflicting results may be complicated. (i) The incidence of toxicities between cisplatin, carboplatin, and oxaliplatin varied . (ii) AlTill date, the candidate gene approach is the most widely used strategy to identify platinum-induced toxicities and their associated polymorphisms. Novel variants will not be identified using this method, although the likelihood of a positive or negative association with a particular adverse event may be considerable in robust studies. Nevertheless, although previous GWAS and whole-exon sequencing had found novel genetic factors associated with platinum-induced toxicities, all these identified interactions need further verification to determine the mechanism.The current meta-analysis has several limitations. First, studies without detailed data were excluded because of the lack of information regarding a significant association between gene polymorphisms and platinum-induced toxicities, and we were unable to contact these authors to provide us with the detailed data. This may have caused publication bias. Second, some of the included studies had small sample sizes, which may result in a decreased power to detect significant differences in the distribution of genotypes between grade 3–4 toxicities and grade 0–2 toxicities. Third, the occurrence of platinum-induced toxicities is affected by various factors, and hence, the pooled OR in this meta-analysis was based on the crude OR from the original studies. Because we could not obtain the raw data from individual studies, the pooled OR in this study was not adjusted for potential confounding factors such as sex, age, smoking status, comedications, gene–gene interactions, and gene–environment interactions, among other factors.The results of our study indicated that single grade 3–4 toxicities that were associated with genetic polymorphisms might partly be the cause for inter-individual differences. Hence, further pharmacogenomics research is needed to determine the novel mutations and their associations. The combined effect of the genetic and clinical factors via gene–gene and gene–environment interactions should be considered in future studies, which may help to predict the risk of lung cancer. Nevertheless, more number of prospective, high-quality, multicenter clinical trials are urgently needed to explore the impact of gene polymorphisms on platinum-based toxicities in patients with lung cancer.MTHFR A1298C and C677T polymorphisms and MDM2 rs1470383 polymorphisms were significantly correlated with platinum-induced severe toxicities in patients with lung cancer. These polymorphisms should be considered for personalized chemotherapy treatment for lung cancer in the future.Our study found that The raw data supporting the conclusions of this article will be made available by the authors, without undue reservation, to any qualified researcher.ZL and WL conceived and designed the study. ZL and YW performed research. ZL and JL conducted data analysis. ZL, HY, and WL accomplishment the manuscript, reference collection, data management, statistical analyses, paper writing, and study design.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest. The reviewer J-YY declared a shared affiliation, with no collaboration, with the authors to the handling editor at the time of review.


Lactobacillus and Bifidobacterium genera. The other strains were Streptococci, Enterococci, and Pediococci. The daily probiotic dose varied between 106 and 1010 colony-forming units (CFU)/gram. Probiotics/synbiotics counteracted CMR factors (endpoint data on BMI: standardized mean difference (SMD) = −0.156, p = 0.006 and difference in means (DM) = −0.45, p = 0.00 and on WC: SMD = −0.147, p = 0.05 and DM = −1.21, p = 0.02; change scores on WC: SMD = −0.166, p = 0.04 and DM = −1.35, p = 0.03) in healthy persons. Overweight/obese healthy people might additionally benefit from reducing total cholesterol concentration . Poor quality of probiotic-related trials make systematic reviews and meta-analyses difficult to conduct and draw definite conclusions. “Gold standard” methodology in probiotic studies awaits further development.We aimed to systematically review the effectiveness of probiotic/synbiotic formulations to counteract cardiometabolic risk (CMR) in healthy people not receiving adjunctive medication. The systematic search (PubMed/MEDLINE/Embase) until 1 August 2019 was performed for randomized controlled trials in >20 adult patients. Random-effect meta-analysis subgroup and meta-regression analysis of co-primary (haemoglobin A1c (HbA1C), glucose, insulin, body weight, waist circumference (WC), body mass index (BMI), cholesterol, low-density lipoproteins (LDL), high-density lipoproteins (HDL), triglycerides, and blood pressure) and secondary outcomes . We included 61 trials (5422 persons). The mean time of probiotic administration was 67.01 ± 38.72 days. Most of probiotic strains were of  Cardiovascular diseases (CVD) are the most prevalent noncommunicable disorders, with cardiometabolic risk factors (CMRF) including obesity , abnormaMetabolic malfunctions of diverse nature, with epigenetic, hormonal, and infectious factors, are involved in the pathogenesis ,7. IntesThe microbiota communicates with the host via toll-like receptors, nuclear factor-ĸB, and mitogen-activated protein kinase , which wConsidering these facts, metabolic impairment is at least a consequence of gut microbiota alteration. The use of probiotics and synbiotics to counteract metabolic disturbances has been reported. Probiotics are “live microorganisms that, when administered in adequate amounts, confer a health benefit on the host”, which has been confirmed in properly controlled studies . SynbiotA few meta-analyses evaluating the efficacy of probiotics and synbiotics in persons diagnosed with diabetes or hypertension have been published ,21,22. H2).Two independent authors (K.S.Z. and K.B.) searched PubMed/MEDLINE/Embase from database inception until 1 August 2019 for randomized controlled trials (RCTs) comparing probiotics and synbiotics with placebo/no-intervention/physical activity/diet to counteract cardiometabolic malfunctions in healthy people with normal weight or moderate/high-risk obesity  AND (RCT OR random* OR placebo*) AND . In the Embase database, the search string was  AND  AND ’ OR ‘quetelet index’ OR ‘body ban mass’ OR ‘body mass’ OR ‘body mass index’ OR ‘cholesterol’/exp OR ‘cholesterol’ OR ‘low density lipoprotein cholesterol’/exp OR ‘ldl cholesterol’ OR ‘cholesterol, ldl’ OR ‘lipoproteins, ldl cholesterol’ OR ‘low density lipoprotein cholesterol’ OR ‘high density lipoprotein cholesterol’/exp OR ‘hdl cholesterol’ OR ‘cholesterol, hdl’ OR ‘high density lipoprotein cholesterol’ OR ‘lipoproteins, hdl cholesterol’ OR ‘triacylglycerol’/exp OR ‘triacylglycerol’ OR ‘triglyceride’ OR ‘triglycerides’ OR ‘tryglyceride’ OR ‘dyslipidemia’/exp OR ‘dyslipaemia’ OR ‘dyslipemia’ OR ‘dyslipidaemia’ OR ‘dyslipidaemias’ OR ‘dyslipidemia’ OR ‘dyslipidemias’ OR ‘blood pressure’/exp OR ‘blood pressure’ OR ‘blood tension’ OR ‘pressure, blood’ OR ‘vascular pressure’ OR ‘plasminogen activator’/exp OR ‘fibrinogen’/exp OR ‘factor 1’ OR ‘factor i’ OR ‘fibrinogen’ OR ‘human fibrinogen’ OR, OR ‘inflammation’/exp OR ‘acute inflammation’ OR ‘bacterial inflammation’ OR ‘inflammation’ OR ‘inflammation reaction’ OR ‘inflammation response’ OR ‘inflammatory condition’ OR ‘inflammatory lesion’ OR ‘inflammatory process’ OR ‘inflammatory reaction’ OR ‘inflammatory response’ OR ‘inflammatory syndrome’ OR ‘reaction, inflammation’ OR ‘response, inflammatory’ OR ‘serositis’ OR ‘sterile inflammation’) AND .A manual review of reference lists from the most recent reviews followed the electronic search. Inclusion criteria were (i) full-text randomized controlled trial, (ii) populations containing >20 adult , (iii) treatment with pro-/synbiotics for at least 4 weeks, (iv) randomization to probiotic/synbiotic vs. controls , and (v) available meta-analyzable change score/endpoint data on any of the following outcomes: HbA1C OR glucose OR OGTT OR insulin OR weight OR waist circumference OR BMI OR cholesterol OR LDL OR HDL OR triglycerides OR blood pressure OR SBP OR DBP OR uric acid OR Plasminogen activator inhibitor-1 OR fibrinogen OR any outcome related to inflammation/endothelial dysfunction. The exclusion criteria were as follows: (i) intervention with microbial agent and adjunctive medication aiming or known to prevent or counteract metabolic dysregulation, e.g., metformin, and (ii) disease, excluding morbid and super obese persons. Data from more than 2-arm studies were abstracted separately for particular comparators; however, placebos were preferentially selected, and regarding dietary comparators, products contained no lactic acid bacteria .https://automeris.io/WebPlotDigitizer/).We used the standard data extraction sheet according to our previous studies ,24,25. DCo-primary outcomes were the changes within glycosylated haemoglobin A1c (HbA1C), glucose, insulin, Homeostatic Model Assessment of Insulin Resistance (HOMA-IR), body weight, waist circumference (WC), body mass index (BMI), lipid profile , high-density lipoproteins (HDL), and triglycerides), and blood pressure. Secondary outcomes included uric acid, plasminogen activator inhibitor-1, fibrinogen, and any outcome related to inflammation/endothelial dysfunction  and leukocyte count). Additionally, we abstracted all-cause and adverse-events discontinuation.http://www.meta-analysis.com). We explored study heterogeneity using the chi-square test of homogeneity, with p < 0.05 indicating significant heterogeneity. All analyses were two-tailed with alpha = 0.05.We conducted a random-effects  meta-ana2) and trials including overweight and obese persons Group differences in continuous outcomes were analysed as the pooled standardized mean difference (SMD) in either endpoint scores (preferred) or change scores from endpoint to baseline (if endpoint scores were not available) using observed cases (OC). For continuous metabolic outcomes, standardized mean difference (SMD) and, where applicable, differences in means (DM) were calculated. The additional analyses included studies with participants with proper BMI value  models without interaction term using DerSimonian–Laird estimator estimation of the amount of heterogeneity. The test statistics of the individual coefficient  for predictors were based on standard normal distribution (z), and the overall test was based on the chi-square distribution (Q statistics following the chi-square distribution with degrees of freedom representing the number of predictors). Meta-regression variables included (i) number of low ROB assessments, (ii) study duration, (iii) mono- vs. multi-strain probiotic intervention, (iv) sample size , and (v) age of participants (mean). Finally, we evaluated funnel plots and conducted Egger’s regression test  to detecn = 2727) of screened studies were excluded, being duplicates and/or after evaluation on the title/abstract level. Two (n = 2) additional articles were identified via hand search. After exclusion of duplicates between the initial search and hand search results, 88 (n = 88) full-text articles were reviewed. Of those, a total of 27 (n = 27) papers were excluded due to not fitting the inclusion criteria. The primary reasons for exclusion were wrong study aim (n = 10); non-healthy participants (n = 7); no probiotic treatment (n = 5); too few participants (n = 3); too short a study duration (n = 2); unavailability of full texts (n = 2); and another language other than English, German, and Polish (n = 1), yielding 61 (n = 61) studies that were included in the meta-analysis  were included [n = 54) [Lactobacillus and Bifidobacterium genera. The other strains ingested by study participants were from Streptococcus, Enterococcus, and Pediococcus genera. The daily doses varied between 106 and 1010 CFU (colony-forming units). The trials were financed by only industry budgets in 20 (n = 20) [n = 10) [As demonstrated in included ,87,88,89[n = 54) ,86,88,89[n = 54) ,79,80,81[n = 54) ,83,88,89[n = 54) ,58,84,86(n = 20) ,79,83,84[n = 10) ,52,58,60n = 2934, 57.22%). Baseline metabolic parameters of included persons are presented in All studies included healthy subjects (including overweight and obese but excluding morbidly obese persons), with a total of 6820 subjected to randomization and 5422 subjected to analysis. The overall mean age was 44.26 ± 12.87 (range: 21.43–71.9) years. The majority of studied persons were females . The highest score, i.e., 7 low ROB assessments was detected in only one study  and 6 lon = 1256, SMD = −0.156, 95%CI = −0.27 to −0.04, p = 0.006 and DM = −0.45, 95%CI = −0.69 to −0.21, p = 0.00 and WC studies = 8, n = 690, SMD = −0.147, 95%CI = −0.30 to 0.03, p = 0.05 and DM = −1.21, 95%CI = −2.27 to −0.16, p = 0.02. In the case of the meta-analysis using change scores, the following results were obtained: WC studies = 5, n = 711, SMD = −0.166, 95%CI = −0.32 to −0.005, p = 0.04 and DM = −1.35, 95%CI = −2.59 to −2.15, p = 0.03 (p = 0.02). For complete results, see Out of all the metabolic indices that we evaluated, BMI and waist circumference decrease were significantly lower with the probiotic compared to controls. For endpoint data, the results were as follows: BMI studies = 16, p = 0.03 . In one As for the other metabolic indices, we found that probiotic ingestion in clinically healthy subjects did not affect those .2) and abnormal (BMI: >25 kg/m2) range, we were able to demonstrate that probiotic intake significantly affected total cholesterol  in persons with normal BMI value . However, Egger’s test did indicate publication bias : t value = 5.38, p = 0.000006). On the other hand, the analyses on the same parameter but regarding change scores depicted that probiotics significantly lowered the parameter in persons with abnormal BMI only . In this case, no publication bias was detected : t value = 1.64, p = 0.137). At last, we evaluated that WC (change score) was significant also in persons with abnormal BMI . In this case, no publication bias was detected : t value = 1.29, p = 0.265).When conducting analysis by BMI status, i.e., in persons with BMI within normal : ROB  and age , indicating that the predicted effect size decreases with increasing risk of bias and increases with age. The model explained 96% of the heterogeneity; however, the permutation test did not confirm the significance of predictors . In the case of insulin, p values for all predictors were found to be significant : number of low ROB assessments , number of persons analysed , duration of probiotic intervention , and BMI of analysed subjects , indicating that the predicted effect size tended to decrease with ROB and study sample size, whereas with increasing duration and BMI, the effect size tended to be greater. The model did not explain heterogeneity, and the permutation test was nonsignificant. When analysing the triglycerides level, p values for all predictors were also significant : monostrain vs. multistrain probiotics  and BMI , indicating that the predicted effect size tended to be smaller for multistrain formulas whereas the effect size increased with increasing BMI. The model explained 90% of heterogeneity, and permutation tests were significant . Finally, for total cholesterol, p values for all predictor were significant : number of low ROB assessments , number of persons analysed , duration of probiotic intervention , age , and BMI of participants . For BMI, HOMA-IR, LDL, and systolic blood pressure (SBP), p values for all predictors and for the effect sizes calculated for change score data were found to be nonsignificant (Q).For endpoint data (SMD) regarding diastolic blood pressure (DBP), n = 13), bacterial metabolites analyses (n = 9), as well as gut barrier integrity markers  (n = 13) and various blood immune markers (mostly cytokines) (n = 12). In 18 studies ,83,85,902) were excluded from the present analysis. We also decided to exclude studies with adjunct medications with reported efficacy against metabolic dysregulation , BMI  kg m−2, I2 = 57%) and fat percentage  %, I2 = 19%). Similarly to our findings, the effect of probiotics on fat mass was not significant  kg, I2 = 84%) [Lactobacillus amylovorus and Lactobacillus fermentum strains reduced this body fat mass [In past years, many studies revealed that probiotics and synbiotics, through interactions with hosts, could affect nutrient metabolism and energy balance. Our current meta-analysis of 61 clinical trials and 5422 persons exclusively investigated the impact of probiotic and synbiotic interventions to reduce cardiovascular risk factors in otherwise healthy adults. The only factor we decided not to exclude was overweight and obesity, as their prevalence is worldwide and as they impact human’s health . Morbidletformin ). Similaetformin ,94,95 Thetformin . This inetformin . Furtheretformin ,98. The etformin  indicaten et al.  confirme2 = 84%) . Also, a2 = 84%)  demonstrfat mass .Bifidobacterium bifidum, Lactobacillus casei, and Lactobacillus acidophilus were ingested for 12 weeks, significantly decreased the plasma glucose and insulin resistance. In a meta-analysis by Samah et al., [The current meta-analysis did not confirm the efficacy of probiotics administration in reduction of other cardiovascular risk in healthy people. Of note, carbohydrate and lipid metabolism was not significantly affected by this type of intervention. In contrary to diabetic patients, we did not find any effect of probiotic therapy on carbohydrate metabolism. A study by Raygan et al.  which wa et al.,  moderate et al., ,105. As  et al., . Similar et al.,  and by T et al., . At last et al., .p < 0.05) in comparison to controls [Lactobacillus, Lactococcus, and Bifidobacterium. These probiotics increase the production of bile salts from cholesterol in their colonized area and, as a consequence, contribute to reduced risk of coronary heart diseases [In our study, we found that probiotics can decrease the total cholesterol level in persons with increased BMI, but other lipid parameters were not affected by probiotics and synbiotics administration. In Wang et al.’s meta-analysis including 32 randomized controlled trials (1971 participants with various metabolic entities), it was proved that probiotics significantly reduced serum total cholesterol , controls . Similarcontrols  and Shimcontrols  (30 RCTscontrols . Bile sadiseases .11 CFU). In contrast to our study, the authors included studies evaluating people with metabolic syndrome, hypertension, and hypercholesterolemia. As the menopause period is a strong contributor of CVD [The administration of probiotics improved blood pressure in humans, which was confirmed in Khalesi et al.’s meta-analysis including 9 randomized, controlled trials . The conr of CVD , we lookr of CVD . Also, ar of CVD , probiotr of CVD  was founr of CVD  did not Last but not least, we abstracted data related to the influence of probiotic administration on gut microbiota and immunological markers. The most frequently studied variables were (i) the effects of probiotic administration on the composition of the microbiota and (ii) colonization with probiotics. Among microbial metabolites, mostly faecal SCFAs were analyzed. The authors analyzed also markers of gut-barrier integrity—mostly LPS and different cytokines as well as inflammatory markers. CRP measured in few studies can be considered as an inflammatory marker as well as a gut integrity marker. Based on the results obtained, no definite association can be found between the use of probiotics, microbiota changes, modulation of the immune system, and either presence or lack of clinical effects . Of noteSeveral limitations of this meta-analysis need to be underlined. These include (i) a relatively small number of high-quality double-blinded studies comparing probiotic intervention to controls with a wide range within the number of participants preceded by no sample size calculations; (ii) heterogeneous study inclusion criteria , and (iii) various type of strains and duration of probiotic intervention. In studies incorporated into the present meta-analysis, the association between the probiotic effect in relation to supplement dose and treatment duration was not analyzed. At last, most of the studies were financed by the industry and include products combined with different ingredients. These all are confounding factors for probiotic efficacy, which may have resulted in some publication bias as evaluated by Eagerr’s test and funnel plots . ConsequProbiotics may counteract some CMRF  in clinically healthy participants. Overweight/obese persons might benefit from the reduction of total cholesterol serum concentration. Poor quality of probiotic-related trials make systematic reviews and meta-analyses difficult to conduct and draw exact conclusions. “Gold standard” methodology in probiotic studies awaits further development.