Neurological diseases are difficult to diagnose in time, and there is currently a lack of effective predictive methods. Previous studies have indicated that a variety of neurological diseases cause changes in the gut microbiota. Alpha diversity is a major indicator to describe the diversity of the gut microbiota. At present, the relationship between neurological diseases and the alpha diversity of the gut microbiota remains unclear.We performed a systematic literature search of Pubmed and Bioproject databases up to January 2021. Six indices were used to measure alpha diversity, including community richness , community diversity , and phylogenetic diversity (PD). Random-effects meta-analyses on the standardized mean difference (SMD) were carried out on the alpha diversity indices. Subgroup analyses were performed to explore the sources of interstudy heterogeneity. Meta-analysis was performed on articles by matching the age, sex, and body mass index (BMI) of the disease group with the control group. Meanwhile, subgroup analysis was performed to control the variability of the sequencing region, platform, geographical region, instrument, and diseases. The area under the curve (AUC) value of the receiver operating characteristic (ROC) curve was calculated to assess the prediction effectiveness of the microbial alpha diversity indices.n = 1,469; controls, n = 1,289). The pooled estimate demonstrated that there was no significant difference in the alpha diversity between patients and controls (P < 0.05). Alpha diversity decreased only in Parkinson's disease patients, while it increased in anorexia nervosa patients compared to controls. After adjusting for age, sex, BMI, and geographical region, none of the alpha diversity was associated with neurological diseases. In terms of Illumina HiSeq 2000 and the V3-V5 sequencing region, the results showed that alpha diversity increased significantly in comparison with the controls, while decreased in Illumina HiSeq 2500. ROC curves suggested that alpha diversity could be used as a biomarker to predict the AD , MS , schizophrenia .We conducted a meta-analysis of 24 published studies on 16S rRNA gene amplified sequencing of the gut microbiota and neurological diseases from the Pubmed and Bioproject database (patients, Our review summarized the relationship between alpha diversity of the gut microbiota and neurological diseases. The alpha diversity of gut microbiota could be a promising predictor for AD, schizophrenia, and MS, but not for all neurological diseases. Neurological diseases are neurological or psychiatric complications resulting from structural or functional abnormalities of the brain or spinal cord caused by various factors, which can be divided into two categories: neurodegeneration and neuropsychiatric diseases. Neurodegeneration includes Alzheimer's disease (AD), Parkinson's disease, Huntington's disease (HD), multiple sclerosis (MS), amyotrophic lateral sclerosis (ALS), motor neuron disease (MND), hepatic encephalopathy, neuromyelitis optica, chronic fatigue syndrome (CFS), Guillain-Barre syndrome, epilepsy, and cerebral ischemia. The other includes depressive disorder, post-traumatic stress disorder (PTSD), attention deficit hyperactivity disorder (ADHD), autism spectrum disorder (ASD), anorexia nervosa, schizophrenia, and anxiety. Neurological diseases are a type of disease with complex etiology, easy recurrence, and difficult timely diagnosis. Their prevalence in developing countries is at least twice that in developed countries , community diversity , and PD.https://pubmed.ncbi.nlm.nih.gov) and Bioproject database (https://www.ncbi.nlm.nih.gov/bioproject), up to January 2021, by using the following retrieval type “Alzheimer's disease *[Title/Abstract] AND  *[Title/Abstract] AND (microbe or microbiota or flora or microorganism or microbiota) *[Title/Abstract]”. In order to more comprehensively search the relevant literature, the disease names in the above search formula were replaced successively by “Parkinson's disease”, “Huntington's disease”, “multiple sclerosis”, “amyotrophic lateral sclerosis”, “hepatic encephalopathy”, “motor neuron disease”, “chronic fatigue syndrome”, “Guillain-Barre syndrome”, “epilepsy”, “cerebral ischemia”, “depressive disorder”, “posttraumatic stress disorder”, “attention deficit hyperactivity disorder”, “autism spectrum disorder”, “anorexia nervosa”, “schizophrenia”, “anxiety”, and “neuromyelitis optica”.We performed a systematic literature search in the Pubmed database  Unpublished data and studies that couldn't distinguish between patient and control groups. PRISMA flow diagram depicts the screening process from the Pubmed and Bioproject database for inclusion of studies  Population study; 2) The source of the sample is a stool or rectal swab; (3) Using 16S rRNA gene amplified sequencing method; (4) The sample size cannot be less than 10 of each group; (5) Patient and control groups can be distinguished; (6) Availability of gut microbiota data. The exclusion criteria were: (1) Editorial, letter, comment, reviews, and meta-analysis; (2) Non gut microbiota-related studies and other diseases; (3) Animal studies or  studies , 2. The sourvia q2-alignment) and used to construct a phylogeny with fasttree2 . The amplicon sequence variants were annotated using the SilvA-138 99% OTU reference sequence pre-trained with the classify-Sklearn Naive Bayes classifier  were executed for alpha diversity indices when each study had statistical heterogeneity . Otherwise, the fixed-effect model was adopted. To explore sources of interstudy heterogeneity, the subgroup analysis controlled for age, sex, body mass index (BMI), sequencing region, geographical region, platform, instrument, and diseases. The area under the curve (AUC) value of the receiver operating characteristic (ROC) curve was calculated to assess the prediction effectiveness of the microbial alpha diversity indices. P < 0.05 was considered statistically significant.As primary outcomes of interest, we extracted the alpha diversity of the gut microbiota at the OTU level. Meta-analysis was performed on the included studies by using Review Manager 5.3 software. For continuous variables, mean and standard deviation (SD) were used as analytical statistics, and each effect size was represented by a 95% CI value. The heterogeneity of the included studies was tested by the A total of 3,600 records were retrieved from the PubMed database, according to the screening criteria, 21 articles  samples were included in the meta-analysis. Twenty four studies compared patients with healthy controls, and 6 compared one or more subgroups with diverse clinical statuses. A total of 22 studies were conducted in Asia , 4 in Europe , 1 in South America, and 5 in North America (the United States of America and Ecuador). In addition, 22 studies used 16S rRNA gene amplified sequencing with the Illumina Miseq platform, 3 studies used Illumina Hiseq 2500, 2 studies used Illumina Hiseq 2000 sequences, and the remaining studies used Roche 454 pyrosequencing Illumina iSeq 100 and NextSeq 500.Neurological diseases, AD, epilepsy, hepatic encephalopathy, ASD, MS, Parkinson's disease, anorexia nervosa, optic neuromyelitis, and schizophrenia were included in the study. A total of 2,758 subjects , Chao1 , and ACE , Shannon , Simpson , PD  , community diversity , and PD. Before controlling for other confounders, the pooled estimate demonstrated that there is no significant difference between AD, epilepsy, hepatic encephalopathy, ASD, MS, Parkinson's disease, anorexia nervosa, optic neuromyelitis, and schizophrenia patients and controls, including observed species  see .P = 0.0001, I2 = 22%)  (P = 0.03) (P = 0.03) (P > 0.05)  , SimpsonI2 = 0%) . However = 0.03)  and Simp = 0.03)  increase.05) see .Meta-analysis was performed on articles matching age, sex, and BMI between the patient and the control group. And subgroup analyses were performed to explore the sources of interstudy heterogeneity. However, the pooled estimate demonstrated that there is no significant difference between these two groups. When the analysis was limited to sequencing platform and geographical region, the result still showed no significant difference.P = 0.98, I2 = 67%)         see  and Simp66%) see . Within I2 = 0%) , 6B. ButI2 = 0%) , Chao1 (I2 = 0%) , ShannonI2 = 0%) , SimpsonI2 = 0%) , and PD I2 = 0%) .P = 0.79, I2 = 76%)    see  and Simp71%) see . But whe = 0.02) .P = 0.001; Simpson, AUC = 0.769, P = 0.0001), MS , schizophrenia   , but not= 0.002) .So far, there has not been a systematic review has comprehensively revealed the association between the gut microbiota and neurological diseases, and previous studies have been limited to individual diseases  and basic characteristics of the disease  related to the subjects in all included studies. Some medications like antidepressants had been shown to have antibacterial effects , National Natural Science Foundation of China Youth Science Foundation project , and Natural Science Foundation of Guangxi (2018GXNSFAA050099).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


It is known that cardiovascular events (CVE) occur more often in winter than in summer. However, dependence of myocardial infarction (MI) risk of on various meteorological factors is still not fully understood. Also, the dependence of the seasonal dynamics of MI on gender and age has not yet been studied. The purpose of our meta-analysis is to reveal dependence of the circannual dynamics of MI hospitalizations on gender, age, and characteristics of a region’s climate.Using Review Manager 5.3, we performed a meta-analysis of 26 publications on the seasonal dynamics of MI. In our meta-analysis, the relative MI risk was higher in colder compared to warmer seasons. Old age insignificantly increased the seasonal MI risk; gender did not affect the seasonal dynamics of MI, but MI was more common in men than in women. The severity of the seasonal dynamics of MI risk depended on the climate of the region. In a climate with a small amplitude of circannual fluctuations in air temperature, atmospheric pressure, and partial oxygen density in the air, as well as in regions where air humidity is higher in winter than in summer, an increase in MI risk in winter compared to summer was significant. It was not significant in regions with opposite climatic tendencies.Based on the results of our studies, it can be concluded that a decrease in air temperature increases in MI risk; in addition, hypoxia in the hot season can provoke CVE associated with ischemia.The online version contains supplementary material available at 10.1186/s43044-022-00322-5. In 2019, approximately 18.6 million people died from cardiovascular diseases, of which 9.14 million deaths were due to ischemic heart disease –5.The study  has demoThe study by Cannistraci et al.  showed tIt is known that MI occurs more often in men than in women, and MI risk increases with age . In addiThe purpose of our meta-analysis is to reveal dependence of the circannual dynamics of MI hospitalizations on gender, age and characteristics of a region’s climate. Elucidation of the relationship between MI risk and fluctuations in weather conditions is an important task for the prevention of MI and mortality from them.A systematic review was conducted following the Meta-analysis of Observational Studies in Epidemiology guidelines . The seaThe search strategies were based on combinations of keywords related to climate  and MI  and hospitalization  and gender/sex  and age . The “Human” filter has been applied. The search was carried out in the databases of PubMed, Scopus, Russian database eLibrary, Google Scholar with a limited publication period of 1940–2021 yy due to imperfections in the diagnostic methods in early studies. The search was carried out in English and Russian. We also searched the reference lists of the included studies.We selected publications on the seasonal dynamics of hospitalizations for MI, but no deaths from MI. The method of displaying statistical data on the paper was supposed to enable us to conduct a meta-analysis. For example, articles in which the number of hospitalizations was presented per 100,000 population or week, or was expressed only as odds ratio, or as regression without indicating absolute values were excluded. In the selected studies for this meta-analysis, data for four seasons  had to be presented. Regions with mountainous climate were not included in the meta-analysis. If a group of authors published several articles on the same topic, the publication with data presented for the maximum observation period was selected.Only high- and medium-quality studies were selected for meta-analysis. When assessing the quality of publications, the following were taken into account: diagnostic methods, duration of observation, study localization, presentation of seasons, presentation of meteorological data, sample size, gender / age representation and presentation of results.Two researchers independently searched for data. After reconciliation, the data were included in the meta-analysis. When conducting a meta-analysis, data of MI cases per season and the total number MI was extracted or calculated from publications. The data were processed without regard to gender and age (data for gender and age groups were pooled), as well as taking into account gender and age. To study the seasonal dynamics of MI risk by gender and age, only those publications were selected in which data were presented separately for gender / age groups. 2)), as described in the work [If the study accurately indicated the localization of the study region, then the publication was included in the meta-analysis, which assessed dependence of MI risk on the characteristics of the climate of the region. If the article presented meteorological data, then we used them. Otherwise, using the archived data, we calculated the meteorological parameters  test was applied to assess MI risk. Study heterogeneity included in the meta-analysis was determined using criterion In et al. . A Z-tesA total of 1279 publications were found. We selected 26 publications reporting the seasonal dynamics of MI , 20–40  do not exceed 16 g/m3; in Seoul, Takashima, Moscow, Chicago, they are more than 28 g/m3. Atmospheric pressure is usually higher and more variable in winter than in summer. For the climate of the temperate zone of Europe, pronounced seasonal dynamics of atmospheric pressure variability, but not its average value, are characteristic. On the contrary, in the regions of East Asia and the Middle East, atmospheric pressure is 10–20 hPa higher in winter than in summer, while the seasonal dynamics of atmospheric pressure variability is weakly expressed. Relative humidity usually varies from winter to summer. In Europe, relative humidity is higher in winter than in summer, while in East Asia, its annual trend is reversed.We studied the seasonal dynamics of MI in 20 cities. Cities were located in various climatic zones, ranging from temperate to tropical (Table p < 0.05). Of all MI cases per year, accounting for winter 26.3 ± 2.7%, summer 23.7 ± 2.5%, spring 25.4 ± 2.6%, and autumn 24.7 ± 2.3%. The relative risk of MI was higher in the colder season than in the warmer one  than in women 33%), in old people (≥ 65 years old) than in young and middle-aged people  can be provoked by fluctuations in blood pressure and blockage of the vessel by thrombus or cholesterol plaque. Previously, it was shown that the seasonal dynamics of IS depend on the climate of the region. A drop in atmospheric pressure and high relative humidity during hot weather were associated with an increased IS risk in summer compared to winter. At the same time, weakly pronounced seasonal dynamics of the average monthly atmospheric pressure and high air humidity in winter are associated with a slight increase in IS risk in winter . It is k2 , the par2 , and the2 , 54. MetS and MI –58. On tS and MI , 60. Proarameter , 5.Some authors have associated an increase in CVE risk in hot weather with an increase in blood viscosity during hyperthermia , 62. It 2, which decreases even more by heat and high humidity. It should be noted that high relative humidity in winter can increase the feeling of cold and promote CVE risk.It is known that atmospheric pressure drop is also associated with the risk of thrombus formation. A 10 hPa decrease in atmospheric pressure has been shown to increase the relative risk of deep vein thrombosis by 2.1% . There iThe regularities of the effect of circannual fluctuations in atmospheric pressure on the seasonal dynamics of CVE associated with ischemia that we discovered allow us to explain the contradictions between the results of clinical observations carried out in different climatic zones. For example, in Oita and Seoul, where in summer compared to winter, atmospheric pressure drops significantly and air humidity increases, heat provokes MI , 35. At In our meta-analysis, the relative MI risk was higher in colder compared to warmer seasons. Based on the literature data, it is obvious that the seasonal MI risk is associated with circannual changes in biochemical and physiological parameters Table . Old ageThe limitation of the meta-analysis that affects the final result may be such factors as different length of the observation period in different publications, different sample size when comparing subgroups, high heterogeneity of results. In addition, in the absence of meteorological data regarding the period of a study, we considered a period of 10 years. This may have led to insufficient accuracy in the processing of meteorological data in such cases. However, it is known that seasonal trends in meteorological factors are quite stable within climatic zones. In this meta-analysis, when studying MI risk from a climate of a region, we compared only two seasons, winter and summer, since the meteorological conditions in these seasons are as stable and contrasting as possible. It is rather difficult to take into account the peculiarities of weather conditions in spring and autumn in a meta-analysis, but it is possible in clinical observations.We did not take into account lifestyle factors  and outbreaks of respiratory viral infections, which may be important for MI risk , 80, 81.Additional file 1. Supplemental material including figures S1–S2 and tables S1–S2.


The current systematic review and meta-analysis was done to evaluate the effects of selenium and probiotic co-supplementation on lipid profile and glycemia indices of the adult population using randomized controlled clinical trials (RCTs).P < 0.001), insulin , homeostatic model assessment for insulin resistance , quantitative insulin sensitivity check index , total cholesterol , low-density lipoprotein cholesterol , and triglyceride .Five studies involving 282 participants with a sample size ranging from 38 to 79 were eligible to be enrolled in the current study. Co-supplementation with probiotic and selenium reduced fasting plasma glucose (WMD =  −4.02 mg/dL; 95% CI: −5.87 to −2.18; The findings of the current systematic review and meta-analysis suggested that co-supplementation with probiotics and selenium may benefit adults in terms of glycemia indices and lipid profile. However, due to the small number of included studies, further trials are needed to confirm our findings.The online version contains supplementary material available at 10.1007/s13668-023-00448-1. With nearly one in three deaths, cardiovascular disease (CVD) is most common cause of death in the USA . Most ofIt has been reported that gut microbiota, microbes living in the human intestinal tract, may affect CVD pathogenesis and related risk factors . Early sProbiotics are defined as living microorganisms that exert beneficial health effects when consumed in adequate amounts. It has been suggested that joint selenium and probiotic supplementation are much more effective than single selenium or probiotic supplementation in terms of metabolic profile •. MoreovSelected electronic databases including ISI Web of Science, PubMed, and Scopus were searched systematically from the earliest available date to February 2022 to find relevant studies. Two independent reviewers conducted a database search to identify studies that investigated the effects of selenium and probiotic co-supplementation on lipid profile and glycemia indices using the following keywords: probiotic OR probiotics OR Lactobacillus OR Bifidobacterium OR Streptococcus OR Saccharomyces OR Enterococcus AND selenium Table . The refThe PICOS  framework was used during study selection as follows: P , I (probiotic + selenium supplement), C (placebo), O (lipid profile  and glycemia indices ), S (RCTs). All the search results were exported to the EndNote X7 software  to be screened by two independent investigators for eligible studies. Inclusion criteria were as follow: original peer-reviewed full-text RCTs recruited > 18 years old subjects with either parallel or cross-over design that implemented co-supplementation of selenium and probiotic and assessed at least one of the outcomes of interest. Exclusion criteria were as follows: non-human studies; recruited < 18 years old individuals; or non-original full-length studies .Eligible articles were screened by two independent reviewers for extraction of the data of interest using pre-defined Excel sheets. The extracted data were as follows: first author’s name, year of publication, study location, characteristics of the study population , study duration, RCT design, the dose of selenium and probiotics, number of probiotic bacteria, and mean and standard deviation (SD) of change in each outcome in the intervention and the control group.The risk of bias of the included studies was examined using the Cochrane Collaboration’s tool . It consI2) index and values equal to 25, 50, or 75% were interpreted as low, moderate, or high heterogeneity, respectively. Visual inspection of funnel plots in combination with Begg’s and Egger’s tests was implemented to assess publication bias. The influence of each study on overall meta-analysis findings was examined via sensitivity analysis. Each time, one study was removed and a meta-analysis was done with the remaining articles to evaluate the robustness of the findings. P values < 0.05 considered statistically significant.Statistical analysis was done using STATA software . For each outcome, data was collected as mean ± SD in a similar unit to estimate the pooled effect size. Weighted mean differences (WMDs) with corresponding 95% confidence intervals (CIs) were calculated for each studied outcome using inverse-variance fixed-effect models. Between effect sizes heterogeneity was estimated using the I-squared  with evidence of a significant heterogeneity . The overall finding was not sensitive to any individual study. No evidence of publication bias was observed  . There was evidence of significant heterogeneity among the included studies . The overall result was not changed following sensitivity analysis. No evidence of publication bias was observed   with evidence of moderate heterogeneity . The overall result was not sensitive to any individual study. No evidence of publication bias was observed   with evidence of a considerable heterogeneity . The overall finding was not sensitive to any individual study. No evidence of publication bias was observed   Fig. .Fig. 5FoP < 0.001) with no evidence of heterogeneity . Sensitivity analysis revealed that no study can influence overall meta-analysis findings. No evidence of publication bias was observed   Fig. .Fig. 6FoP = 0.029) with no evidence of heterogeneity . The overall finding was sensitive to Raygan et al.  and Shabani et al.  studies. No evidence of publication bias was observed   with no evidence of significant heterogeneity . The overall result was not changed following sensitivity analysis. No evidence of publication bias was observed  . There was no evidence of significant heterogeneity among the included studies . The omission of each study did not change the overall meta-analysis finding. No evidence of publication bias was observed  , insulin (= −2.50 mIU/mL), HOMA-IR (−0.59), and increased QUICKI (0.01). It was reported among Asia Pacific region that each 18 mg/dL lower FPG was associated with a 23% lower risk of ischemic heart disease and a 21% lower risk of total stroke . A meta-Results of our meta-analysis revealed a significant reduction in TC (−12.75 mg/dL), LDL-C (−7.09 mg/dL), and TG −14.38 mg/dL) but not in HDL-C (0.55 mg/dL) following selenium and probiotic co-supplementation. A meta-analysis reported that a 1 mmol/l (38.67 mg/dL) reduction in LDL-C level linked with a 19% reduction in CVD-related mortality and a 12% reduction in all-cause mortality  mg/dL bu–41 regarn = 5) may diminish the precision of pooled effect estimates. Substantial heterogeneity should also be considered that can reduce the generalizability of our findings. Participants’ characteristics , sample size, study duration, dose, and strain of probiotics are among the possible sources of heterogeneity. However, due to the small number of the included studies, we were unable to run a sub-group analysis to find the sources of heterogeneity.The strength of our work is that this systematic review and meta-analysis is the first study that comprehensively pooled the results of available literature regarding the role of probiotics and selenium co-supplementation on lipid profile and glycemia indices. Our findings have research and clinical implications and can be used by clinicians and health practitioners. However, some limitations should be considered while interpreting the results. The small number of the included studies Below is the link to the electronic supplementary material.


Several published studies have examined the association of coffee consumption with atrial fibrillation (AF) risk, but their findings are still controversial. Therefore, we performed a systematic review and dose–response meta-analysis of prospective studies to determine the relationship between coffee consumption and the risk of incident AF.We systematically retrieved the PubMed and Embase databases until October 2021 for pertinent studies that reported the association of coffee consumption (caffeinated or decaffeinated coffee) with AF risk. A cubic spline random-effects model was used to fit the potential dose–response curve. The effect estimates were expressed as adjusted risk ratios (RRs) and 95% CIs.Pnon–linearity = 0.25). Compared with participants with no coffee consumption, the RRs (95% CI) of AF risk estimated directly from the dose–response curve were 1.01 (0.98–1.03), 1.00 (0.97–1.04), 0.99 (0.92–1.02), 0.95 (0.89–1.01), 0.94 (0.87–1.01), 0.89 (0.79–1.02), and 0.87 (0.76–1.02) for 1–7 cups of coffee per day, respectively. One cup per day increased in coffee consumption was associated with a 2% reduced risk of AF .A total of 10 prospective studies (11 cohorts) involving 30,169 AF events and 723,825 participants were included. In the dose–response analysis, there was a linear inverse association between coffee intake and risk of AF although not statistically significant (Our evidence from this meta-analysis suggested that coffee consumption had a trend toward reducing the risk of AF in a dose–response manner. Further studies could be conducted to reinforce our findings. Atrial fibrillation (AF) is the most common cardiac arrhythmia, with the incident rate increasing consistently over the years . The preHowever, the dose effects of coffee intake on the risk of AF remain unclear. Due to the excitatory effect of caffeine on sympathetic nerves, coffee or caffeinated products are considered a trigger factor for supraventricular arrhythmias in the 2003 American College of Cardiology/American Heart Association Task Force on Practice Guidelines and the European Society of Cardiology (ACC/AHA/ESC) guidelines for the management of patients with supraventricular arrhythmias . SubsequThis meta-analysis was carried out based on the Cochrane Handbook for systemic reviews. The results of this study were reported according to the Preferred Reporting Items for Systematic Reviews and Meta-Analyses (PRISMA) 2020 Statement . EthicalIn total, two reviewers (X-L and WG-Z) systematically retrieved relevant studies on the association of coffee consumption with the risk of AF from the PubMed and Embase electronic databases until October 31, 2021. The search terms included two parts: (1) related to the participants of “atrial fibrillation”; and (2) related to the risk factor of “coffee.” These two categories of keywords were combined using the Boolean operator “and.” The search strategies in the electronic databases are shown in We applied the following inclusion criteria during the study selection. Prospective cohort or nested case–control studies were included if they reported the association of total coffee consumption (caffeinated or decaffeinated coffee) with the risk of AF and/or atrial flutter. There were no strict restrictions on the studied population. Coffee consumption was analyzed as a continuous or categorical variable, and the association with AF risk was expressed as adjusted effect estimates. The assessment methods for AF or atrial flutter were applied according to the originally included studies.Studies were excluded if they only assessed the relationship between caffeine consumption and AF development. The outcome of postoperative AF was not included. We also excluded certain publication types such as retrospective or cross-sectional studies, case reports, editorials, meeting abstracts, and comments. If facing with overlapping data, we selected the study with the largest sample size or the longest follow-up duration.The PRISMA flow diagram was applied to guide the process of study selection. Studies were selected by two reviewers independently based on the eligibility criteria mentioned above. We first screened the titles and abstracts for the potential studies, and then selected the final included studies after the full-text screenings. Disagreements were resolved through discussion or consultation with a third researcher.Data from each included study were extracted by two reviewers independently, which were further confirmed by a third researcher to ensure accuracy. We extracted information including the study characteristics , study participants , coffee information , ascertainment of AF, adjusted factors in the multivariable analysis, follow-up period, and outcome data . If the included studies reported adjusted effect estimates in multiple models or different follow-up periods, we applied the most adjusted one or the effect estimate in the longest follow-up period.The quality assessment was performed using the Newcastle-Ottawa Scale (NOS) tool. This tool had three domains ranging from 0 to 9 points, including the selection of cohorts (0–4 points), the comparability of cohorts (0–2 points), and the assessment of the outcome (0–3 points). According to the previous studies , 17, theP-value of the Cochrane Q test and the I2 value, where P < 0.1 or I2 > 50% suggested significant between-study heterogeneity. If facing an I2 value of >50%, we excluded each included study at a time to find out the potential source of high heterogeneity. We assessed the potential risk of publication bias by visually inspecting the funnel plots. In addition, the statistical publication bias was further assessed using the Egger’s and Begg’s tests, where a P-value of >0.1 indicated no significant publication bias.The statistical heterogeneity across the included studies was assessed using the Both semiparametric and parametric methods were applied to assess the effect of coffee consumption on the risk of AF. In the semiparametric methods, the categorizations of coffee consumption included three parts, namely, lowest, second highest, and highest coffee intake levels according to the reports on coffee consumption and risk of diabetes . The assIn the sensitivity analysis, re-analysis with a fixed-effects model was applied. In addition, we also re-performed the analysis after excluding the studies with data on decaffeinated coffee, or the healthy population. In addition, the included studies with adjustment for age, sex, ethnicity, education, smoking, alcohol intake, body mass index, physical activity, diabetes, hypertension, coronary heart disease, or heart failure were separately pooled. The subgroup analysis was performed based on sex .1), and the Stata software . In this study, a P < 0.05 indicated statistical significance.All the statistical analyses were performed using the Review Manager version 5.4 software  and Embase (n = 135) databases. We found no additional studies through the reference lists of prior meta-analyses and reviews  nor the second-highest (median 2.5 cups/d) vs. the lowest coffee intake level  was significantly associated with an increased risk of AF  , 27–31 wsk of AF . We failThe sensitivity analysis obtained similar results using a fixed-effects model . As showP = 0.02). In the sensitivity analysis, a fixed-effects model analysis generated similar results  , 27–31 w results . Per cup results .Pnon–linearity = 0.25). Compared with participants with no coffee consumption, the risk of AF estimated directly from the dose–response curve for 1–7 cups/day were 1.01 (95% CI 0.98–1.03), 1.00 (0.97–1.04), 0.99 (0.92–1.02), 0.95 (0.89–1.01), 0.94 (0.87–1.01), 0.89 (0.79–1.02), and 0.87 (0.76–1.02), respectively.As shown in Pinteraction = 0.08; Pinteraction = 0.42; Pinteraction = 0.48; In the categorical analysis of the association of coffee intake with AF risk, there were no interactions between sex regarding the association of the highest  neither the highest (median ≥ 4 cups/d) nor the second-highest (median 2.5 cups/d) coffee intake level compared with the lowest level was associated with an increased risk of AF; (2) one cup per day increase in coffee consumption was with a decreased risk of AF. There was a linear inverse relationship between coffee consumption and AF risk although not significant, suggesting that AF risk tended to decrease with higher intakes. Overall, our current evidence indicated that moderate to high levels of coffee consumption at least did not increase the incidence of AF.A prior meta-analysis conducted by Krittanawong et al.  includedAs a complex beverage roasted from plants, coffee contains more than 1,000 biologically active chemicals, including caffeine, chlorogenic acids, diterpenes, cafestol, kahweol, trigonelline, melanoidins, and modest amounts of electrolytes and vitamins . Some ofIn our pooled analysis, we combined caffeine and decaffeinated coffee as total coffee due to the lack of an exact number of participants consuming either kind of coffee in every included study. Additional data from four cohort studies reporting caffeinated coffee still suggested no positive associations between coffee consumption with AF risk. Decaffeinated coffee was not analyzed in our study due to limiting data. However, different brewing methods would make the coffee components differ a lot , 34, whiRecently, Marcus et al. designed an n-of-1 randomized clinical trial and found that except for alcohol, other self-selected triggers  are neither related to the AF-related quality of life nor the onset of AF . The n-oOur analysis had several limitations that should be carefully addressed. First, because caffeinated and decaffeinated coffee may have different effects on AF risk, future studies should examine subgroups based on different types of coffee separately. Second, in the studies we included, coffee consumption was calculated based on individual reports, and they did not explicitly state the brewing methods of coffee. Therefore, the exact volume could not be accurately obtained, nor did the content of various ingredients in the coffee. Third, the studies included in this meta-analysis were mainly conducted in the United States and European countries. Future research in different populations would help draw conclusions about the relationship between coffee consumption and the risk of AF worldwide. Finally, no strict restrictions on the populations were applied during the inclusions of studies. Nevertheless, we separately pooled 8 cohorts enrolling the healthy population, suggesting similar findings.Our meta-analysis suggested that coffee consumption had a trend in reducing the risk of AF in a dose-response manner. Coffee intake at least did not increase the incidence of AF, conveying to public health that coffee intake was not a trigger of AF development.The original contributions presented in this study are included in the article/All authors listed have made a substantial, direct, and intellectual contribution to the work, and approved it for publication.The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. Any product that may be evaluated in this article, or claim that may be made by its manufacturer, is not guaranteed or endorsed by the publisher.


Humans continue to be constantly exposed to mycotoxins, mainly through oral exposure (dietary), inhalation, or dermal contact. Recently, it has been of increasing interest to investigate mycotoxin-linked carcinogenicity. This systematic review was conducted to synthesize evidence of the association between mycotoxin-linked mutations and the risk of cancer, to provide an overview of the data linking exposure to different mycotoxins with human cancer risk, and to provide an update on current research on the risk of cancer associated with human exposure to mycotoxins. PRISMA guidelines were used when conducting the systematic review. PubMed, MEDLINE, and CINAHL electronic databases were comprehensively searched to extract the relevant studies published from inception to May 2022. A total of sixteen relevant studies (4907 participants) were identified and included in this review. Of these, twelve studies were from Asia, while four of the studies were conducted in Africa. The overall meta-analysis result found no significant association, although some of the studies confirmed an association between mycotoxin-linked mutations and primary liver cancer risk. Mainly, the experimental studies have shown associations between mycotoxin-linked mutations and cancer risk, and there is a need for researchers to confirm these links in epidemiological studies in order to guide public health policies and interventions. Mycotoxins are toxic fungal metabolites which are structurally diverse and cause adverse human and animal health issues, mainly by oral exposure. They are the common contaminants of agricultural food/feed products, such as maize, wheat, nuts, and other foods derived from them. To date, mycotoxins with carcinogenic potency as reported in the literature include aflatoxins, ochratoxin, fumonisins, zearalenone, and some Penicillium toxins. Most of these carcinogenic mycotoxins are genotoxic and mutagenic agents in many model systems and produce chromosomal aberrations. In various climatic conditions, fungi can produce different mycotoxins, while some mycotoxins can be produced by different fungal species . HoweverF. graminearum [A variety of factors influence the fungal and mycotoxin contamination of agricultural commodities. Among them are physical/environmental, chemical, and biological factors . Physicaminearum .Lastly, one of the biological factors influencing mycotoxin production is the interaction between toxigenic fungal species and the substrate. Among the factors are strain specificity, strain variation, and eventually the instability of toxigenic properties ,14. HowePenicillium toxins [Mycotoxins with carcinogenic potency as reported in the literature include aflatoxins, fumonisins, ochratoxin, T2, zearalenone, and some m toxins .Aflatoxin (AF): Aflatoxin B1 is the most carcinogenic among all mycotoxins, with the liver as its predominant target . In humaAccording to Peraica et al.,  fumonisiAspergillus toxins, only OTA is potentially as important as the aflatoxin [Ochratoxin A (OTA): Among the flatoxin . The kidflatoxin . It is aflatoxin . Furtherflatoxin . Howeverflatoxin ,26. Addiflatoxin .Fusarium mycotoxins, which usually contaminate unharvested grains that are left for a long time in the field. T-2 toxin is also caused alimentary toxic aleukia (ATA), which led to the death of many people in Russia. Alimentary toxic aleukia (ATA) is characterized by fever, skin, and nose bleeds, necrosis, and, apart from its cytotoxicity, also suppresses the immune systems of the affected individuals [T-2 toxin is one of the ividuals . T-2 toxividuals .Fusarium mycotoxin. It is a common contaminant of maize but can also affect other crops [Zearalenone (ZEA) is also a er crops . Being aer crops . Due to er crops . Therefoer crops , perhapser crops .The Preferred Reporting Items for Systematic Reviews and Meta-Analyses guidelines (PRISMA) ,33 were This systematic review considered all original studies done on human exposure to different mycotoxins and the associated cancer risk. Both cohort and case control study designs were eligible. The databases searched included: PubMed/MEDLINE, CINAHL, grey literature through Google Scholar, and the reference lists of the papers reviewed. The main Boolean operators  were used to combine search terms in different combinations. Search terms included combinations of: (mycotoxins), , (aflatoxins), (ochratoxin A), (fumonisins), , (neoplasms), (mutations), (cancer), (exposure), (human). The search included articles published from inception up to 31 May 2022, and only articles published in the English language were searched.Eligibility criteria for inclusion in the review were as follows: (1) Studies that investigate the link between any mycotoxin-linked mutations and risks of one or more types of human cancer; (2) Case-control and cohort study designs; (3) Relative risk (RR) or odds ratio (OR) estimates with 95% confidence intervals (CIs) reported, or data to calculate them.Studies were excluded if they were animal studies, drug trials, diagnostic trials, case reports, studies only reporting qualitative findings, interventional studies, or studies not focusing on the association between mycotoxin-linked mutations and risk of one or more type of cancers in humans.Data extraction and management were done independently and in duplicate by two review authors (TI and AM). Titles and abstracts retrieved from various electronic bibliographic databases were first screened for inclusion, and then full texts were examined in detail and screened for eligibility. Reference lists of eligible studies were hand-searched for additional articles. Data were extracted from articles using a template designed for this review. The data extracted from each study included: The first author’s last name, publication year, study population, study designs and sample size, follow-up period, gender, age, number of cases, exposure, outcome measurement, outcome measurement method, biomarkers, outcomes of significance to the review question, and objectives. Odds ratios (OR) and risk ratios (RR), with their corresponding 95% confidence intervals (CIs) reported, and variables adjusted for each measurement outcome in the analysis or data to calculate these if not already calculated, were also extracted.The quality and risk of bias of the selected studies was assessed independently and in duplicate by two review authors (TE and AM) using a modified Newcastle–Ottawa Scale for observational studies  and theyp-value < 0.1 indicating significance) and I-square statistic . Meta-analysis results were displayed using a forest plot.Where results were reported in a similar manner, and where there were no significant differences in results, for instance in the case of primary liver cancer (PLC), odds ratios were pooled using a random effects meta-analysis carried out in Stata . Heterogeneity between results in the meta-analysis was determined using both the Chi-square test  (Results are summarized in  = 9.1%) . For breIn this systematic review, the aim was to synthesize the available evidence from epidemiological studies on the effects of mycotoxin-linked mutations on the risks of cancer. It also provides a summary of the data on the association between mycotoxin exposure and cancer risk in humans. This review found relatively few studies on the subject. The majority of the studies focused on aflatoxin exposure. The sixteen studies included focused on the links between mycotoxin exposure and (1) primary liver cancer risk, (2) breast cancer risk, (3) cervical cancer risk (4) colorectal cancer risks and (5) oesophageal cancer risk. Meta-analysis results found no significant association between aflatoxins and risk of primary liver cancer .Most publications in this study examined the association between aflatoxins and liver cancer, and the relationship between aflatoxin-linked mutations and liver cancer risk was confirmed. The results are in line with the conclusions of the World Cancer Research Fund (WCRF) that there is a strong association between dietary aflatoxin and the risk of liver cancer . FurtherAdditionally, no significant association was found in the two included studies investigating exposure to fumonisin B1 and HCC. However, consistent with IARC classifications, fumonisin B1 can possibly modify protein synthesis. Moreover, the obstruction of DNA synthesis can occur by higher concentrations in intestinal cells in vitro ,68,69. ATwo case–control studies (incorporated into this review) conducted in Africa studied the links between zearalenone and its metabolites, e.g., α-zearalenol in relation to the risk of breast cancer. Although the results were conflicting, it is of note that different biological matrices were used, which cannot easily be compared. One of the studies examined urine  while thZearalenone was shown to be able to initiate cancer in rats, e.g., hepatocellular adenomas and pituitary tumours ,79. ConvAn epidemiological study by Pillay et al., , in SoutA study by Ouhibi et al. , in TuniThe authors found the co-occurrence of the mycotoxins  in the urine samples. While T-2 toxin was only present in the oesophageal cancer group, HT-2 was the most common mycotoxin in combination present in all the co-occurring samples, followed by NEO, which was present in two of the three multi-contaminated urine samples . AlthougMycotoxins are ubiquitously found all over the world, in many foods and feedstuffs, which may be chronically consumed by majority of people all around the world. Most studies investigating the associations between mycotoxin exposures and cancer risk have mainly concentrated on aflatoxins, while other important mycotoxins, for example, ochratoxins, etc., are not well dealt with. Therefore, having known mycotoxins to have toxigenic potentials, there is need for continuous research in order to understand the mechanisms of their carcinogenic effects, also taking into consideration the co-occurrence of different mycotoxins in food and the synergistic effects with other mycotoxins, for public health purposes and the prevention of economic losses.This work has successfully assessed the link between mycotoxin-linked mutations and the risk of cancer. Mycotoxins are a ubiquitous contaminant of many foodstuffs and agricultural products, and the carcinogenic potency of some mycotoxins such as aflatoxins are already known. Thus, inasmuch as the meta-analysis result did not find a significant association, some experimental studies did establish the link between mycotoxin exposures and cancer risk. Additionally, many emerging mycotoxins have not been investigated with respect to their health outcomes. We therefore propose the need for these links to be confirmed and validated using more human epidemiological studies, taking into consideration mycotoxin co-occurrence in food and their synergistic potentials with one another. This will help guide public health policies and interventions.While we did a comprehensive search in the major electronic databases, it is possible that we could have missed some non-English studies. Furthermore, the comparability of the study results was limited because of the different study populations and different biological matrices used for the exposure assessments.


Cognitive impairment is an increasingly urgent global public health challenge. Dietary Inflammatory Index (DII) is a literature-derived score that links diet to inflammation. The relationship between DII and cognitive impairment remains controversial. Therefore, our study aimed to analysis the role of DII on the risk of cognitive impairment by meta-analysis.PubMed, Cochrane Library, MEDLINE, Web of Science and EMBASE databases were searched up to July 2022. Newcastle–Ottawa scale (NOS) and Joanna Briggs Institute (JBI) Checklist were performed to estimate the quality of studies.Nine observational studies with 19,379 subjects were included. Our study found that higher DII could elevate the risk of cognitive impairment . Meanwhile, the OR of cognitive impairment was 1.49  for cross-sectional studies and 1.42  for cohort studies, respectively.Our meta-analysis indicated that higher DII (indicating a more pro-inflammatory diet) is related to increased risk of cognitive impairment. Cognitive impairment is an increasingly urgent global public health challenge, with more than 100 million adults predicted to develop dementia by 2050  effect size for the highest (pro-inflammatory diet) to the lowest (anti-inflammatory diet) DII scores were reported in these studies; (4) occurrence of cognitive impairment as an study outcome; (5) Animal and in vitro studies, duplicate and conference literatures, or reviews were excluded. Two authors assessed all studies independently, resolved disagreements via discussion, and collected final included studies (The inclusion criteria were as follows: (1) observational study including cohort, case–control and cross-sectional design; (2) exposure: inflammatory diet evaluated  studies .All included studies were examined and following data were collected: first author, study area, publication year, study type, numbers, age, gender and cognitive situation of subjects. Meanwhile, the Newcastle–Ottawa scale (NOS) and the Joanna Briggs Institute (JBI) Checklist were used to evaluate the quality for cohort and cross-sectional study, respectively. Moreover, GRADE system was performed to assess quality of the evidence comments.via the software Stata . The I2 statistic was used to assess the statistical heterogeneity and P < 0.05 was defined significant for heterogeneity. In our study, random effects models were used in all analyses based on the level of heterogeneity. Egger's test was performed to evaluate the publication bias, and the trim-and-fill method  was conducted to analysis the influence of bias on the results. Meanwhile, subgroup analyses were performed based on the study area , gender of subjects  and study type .All statistical analysis were performed n = 3) and the JBI are all ≥ 14 for cross-sectional study (n = 6)  . Meanwhi (n = 6) .Our meta-analysis showed that higher DII scores had a 46% elevated risk of cognitive impairment . However, there was no significant difference for results after using trim-and-fill method. Therefore, the effect of publication bias was considered slight and the results were stable . MCI is a transitional state between healthy aging and dementia, characterized by cognitive decline but relatively complete activities of daily living  could increase the risk of cognitive impairment. More data from clinical trials are needed to verify the association.The original contributions presented in the study are included in the article/supplementary material, further inquiries can be directed to the corresponding author.PL and YJ designed the study. PL, YJ, SY, MS, and CW performed the study. YJ and YY analyzed the data and wrote the manuscript. YJ and LW participated in revising the manuscript. All authors agreed with the final version of the manuscript.


Alveolar osteitis (AO) is a common postoperative complication of third molar extractions that is thought to be associated with the intake of oral contraceptives (OCPs). This meta-analysis sought to evaluate the risk of AO associated with OCP use and sex independently and whether this risk was affected by the use of postoperative analgesics or antibiotics.  PubMed/Medline, EMBASE, and Cochrane databases were searched for articles pertaining to OCP use and the incidence of AO using MESH terms. The measured outcome was the development of AO following a third molar extraction. Additional variables such as sex, analgesic, and antibiotic use were documented and included in the analysis. The data were analyzed in R using the Mantel-Haenszel method. p < 0.01). The pooled-RR of AO in females not taking OCPs was not significantly different from males (p=0.45).  Fifteen studies with a total of 1366 female participants who were OCP users and 2919 nonuser female participants were included in this meta-analysis. OCP users were approximately twice  as likely to develop AO following a third molar extraction when compared to nonuser females. The increased incidence of AO in the OCP group was statistically significant ( OCP use significantly elevated the risk of AO in females. Females who did not take OCPs had a similar risk of developing AO compared to males, suggesting that OCP use is a potential effect modifier. Neither postoperative antibiotics use nor the type of postoperative analgesic significantly affected AO incidence in those taking OCPs. Alveolar osteitis (AO) is one of the most common postoperative complications of third molar extractions . AO, refWhile the definition of AO may differ, most would agree that AO involves postoperative pain at the site of dislodgement of the blood clot. The patient often experiences severe pain, halitosis, foul taste, lymphadenopathy, and increased healing time . These sOCPs are thought to increase fibrinolysis by increasing serum plasminogen while decreasing serum plasminogen activator inhibitor-1 . Hence, The primary objective of this systematic review and meta-analysis was to assess the risk of AO in OCP users. The secondary objectives included the analysis of several variables to assess their effect on the incidence of AO in OCP users, including sex, intake of preoperative and postoperative antibiotics, postoperative opioid and nonopioid analgesics, the menstrual cycle, smoking, and the decade of publication of the study. The latter in order to determine if the formulation was a determining factor.This review was planned and conducted in accordance with the Meta-Analysis of Observational Studies in Epidemiology (MOOSE) , and theStudy participants (patient/population): females who took OCPs before, after, or at the time of extraction compared to females who did not take OCPs, all of whom required (intervention) extraction of their third molar(s). No age, race, or health restrictions were applied.Type of Study: Observational Human Studies  and Clinical Trials were selected for the purpose of this study. Case Reports, Case Series, Letters to the Editor, Personal Communications, and Qualitative Studies were excluded.Outcome: AO was clinically diagnosed by postoperative pain and total or partial loss of the blood clot at the site of extraction 1–7 days following surgery.Reporting of results: studies were only eligible if they reported the frequency and a measure of association (Odds Ratio (OR), and/or Relative Risk (RR) with a 95% confidence interval (CI)) with AO cases among OCP users and nonusers (Controls).Length of OCP use (Time): no restrictions applied.Accessibility of data: studies were only eligible if they were published as full publications in the English language.To be considered for this review, all studies were required to meet the following criteria:The literature search was conducted using PubMed/MEDLINE, EMBASE, and Cochrane Library databases, via the following search terms:  OR  AND  OR (dry socket [MeSH Terms]) until March 2022. All search terms were adapted for each database, as necessary. To ensure completeness, the references in the articles selected for inclusion were searched manually and independently by investigators  for potentially relevant articles.Two investigators (MT and DGA) performed independent article searches that met the eligibility criteria. Publications were further examined based on the full title and published abstract. At this point, duplicates and articles that did not meet the eligibility criteria were removed, and a full-text article was obtained for those deemed eligible for inclusion. Once completed, the investigators compared the results, and any discrepancies were resolved by a third investigator (JAS).Two investigators (MT and DGA) independently extracted data using a data abstraction form including studies' characteristics such as the country in which the study was conducted, population characteristics , study design, length of follow-up, exposure (OCP use), time to outcome measure (development of AO), and strength of association. Any discrepancies in the data abstraction were reviewed and resolved by a third investigator (JAS).The meta-analysis was conducted using the “meta” package in R version 4.0.2 . The dat2), where I2 > 30%, I2 > 50%, and I2 > 75% indicated moderate, substantial, and considerable heterogeneity, respectively. Cochran's Q test was used in conjunction with a forest plot to determine the significance (p ≤ 0.05) of the calculated heterogeneity between studies.The degree of dissimilarity in the results of individual studies (or heterogeneity) was assessed graphically using forest plots, and the exact binomial CIs were calculated. Statistical heterogeneity between the reviewed studies was quantified with the DerSimonian and Laird estimate  with the “Quality Assessment Tool for Observational Cohort and Cross-Sectional Studies” of the National Institute of Health (NIH) . All stuThe risk of publication bias was assessed by visual analysis of a funnel plot , where aSubgroup analyses were conducted based on intake of postoperative antibiotics and postoperative opioid or nonopioid analgesics, smoking status, the decade of publication of the study, and menstrual cycle.PubMed/MEDLINE, EMBASE, and Cochrane Libraries identified 1446 articles that met the inclusion criteria, 240 of which were duplicates. 1176 articles were further eliminated after abstract review. The remaining thirty were reviewed by full-text, of which seventeen were deemed eligible for the systematic review and fifteen for the meta-analysis .The included seventeen , 12–27 sAll included studies were of good methodological quality , and thep < 0.01). There was substantial heterogeneity  across the studies . There was substantial heterogeneity across the twelve studies   .n = 2407) were 1.1 times  more likely to develop AO following a third molar extraction when compared to male non-OCP users (n = 4678). This difference was not statistically significant (p=0.45). There was moderate heterogeneity across the twelve studies   .n = 1127) with males (n = 4678), the analysis showed that OCP users were more than two times as likely  to develop AO following third molar extraction. This difference was statistically significant (p < 0.01). There was substantial heterogeneity across the twelve studies   .p=0.02) the chance of developing AO when compared to females who did not take OCPs, while females who took OCPs, nonopioid analgesics, and opioids had a pooled-RR of 1.53 . The difference in the pooled estimate between the two subgroups was not statistically significant (p=0.37). There was substantial heterogeneity among studies   .The female participants of six –22, 25 op=0.04) and approximately twice  the risk of developing AO following a third molar extraction when compared to females who did not take OCPs in regard to postoperative antibiotic use “no” and “yes,” respectively. The difference between the two subgroups (“no” vs. “yes”) was not statistically significant (p=0.88), and there was substantial heterogeneity among studies   .This meta-analysis included fifteen studies and found that OCP use was a significant risk factor for AO. The two clinical trials that were excluded from the meta-analysis also found that female OCP users were more likely to develop AO. In this study, females taking OCPs were approximately twice as likely to develop AO following third molar extractions compared to females not taking OCPs .The female sex is an often-mentioned risk factor for AO. However, the correlation is controversial, and OCP use remains an important effect modifier . The evip=0.41), further suggesting that OCP is a risk factor for AO.The pooled risk for females when controlled for OCP use was not significantly different from the males . Bienek and Filliben [A recent meta-analysis by Bienek and Filliben  examinedFilliben  includedFilliben . FurtherRosendaal et al.  noted thThe pooled-RR of AO was 2.13 times higher in females who were taking OCPs compared to non-OCP users when prescribed postoperative NSAIDs only. Female participants who took both postoperative opioids and NSAIDs had a pooled-RR of 1.53 times higher than non-OCP users .There is no clear consensus on the use of prophylactic antibiotics for the prevention of AO , and theOnly one of the studies  includedp=0.88), suggesting that postoperative antibiotics do not offset the increased incidence of AO associated with OCP use.On the other hand, the pooled-RR of AO was approximately two times higher in those taking OCPs and postoperative antibiotics than in females only taking postoperative antibiotics and in those taking OCPs without postoperative antibiotics than in females taking only postoperative antibiotics . The difHormonal changes, particularly in estrogen, precipitated by the menstrual cycle may have an effect on the incidence of AO. Nordenram and Grave  suggesteThe included studies did not provide sufficient evidence on how OCP use and concurrent smoking are associated with AO. Studies were subgrouped by whether smokers were included in the study or not . The incThis meta-analysis was able to confirm that the intake of oral contraceptives (OCP) is a risk factor for alveolar osteitis (AO) after third molar extraction. According to the CDC, in 2015–2017, 64.9% of the 72.2 million women aged 15–49 in the US were currently using contraception, 12.6% of which were using oral contraceptive pills. We believe that the results of our study are impactful at a predoctoral and postdoctoral level and useful for interdisciplinary and continuing education purposes, as this is something both general dentists and specialists should be aware of. The high prevalence of women using OCPs makes this a relevant finding for the care of our patients.A dose-response relationship may exist between the hormone(s) concentration in OCPs and the incidence of AO, and the menstrual cycle phase the patient is in during the extraction may be a risk factor. This paper will provide other researchers with the premise for potential further research regarding the topic.The difference between females not taking OCPs and males was not significant, suggesting that the female sex is not a risk factor for AO. OCP use was likely an effect modifier in previous studies. Those considering third molar extraction should be advised to temporarily discontinue OCPs prior to surgery. Further investigation should be conducted to evaluate how far prior to surgery OCPs should be discontinued to minimize the risk of AO. Despite the decreasing quantity of estrogen in OCP formulations over time, this meta-analysis did not find evidence that supports the theory that the decreasing quantity of estrogen in OCP formulations would result in a decreased incidence of AO. Neither the type of analgesic taken nor taking antibiotics postoperatively had a significant effect on the incidence of AO. The evidence supporting the menstrual cycle phase as a risk factor for AO was limited.


MTHFR on the risk of autoimmune diseases with inconsistent results. Therefore, this meta-analysis was designed to assess the association between the MTHFR 677 C/T and 1298 A/C polymorphisms and the susceptibility to autoimmune diseases. We identified reports by a literature search in the following electronic databases: PubMed, Ovid, Web of science, and China National Knowledge Infrastructure. Statistical analyses of the summary odds ratios (ORs) and 95% confidence intervals (CIs) were done using STATA software. In a recessive genetic model, the MTHFR 677 C/T polymorphism was associated with an increased risk of Behcet's disease , multiple sclerosis , and ankylosing spondylitis . The MTHFR 1298 A/C polymorphism was associated an increased risk of multiple sclerosis in a heterozygote comparison  and in a dominant model . This meta-analysis demonstrated that the MTHFR 677 C/T was a risk factor for Behcet's disease, multiple sclerosis, and ankylosing spondylitis, and the 1298 A/C was a risk factor for multiple sclerosis.Methylenetetrahydrofolate reductase (MTHFR) is a critical rate-limiting enzyme in the homocysteine/methionine metabolism pathway that is implicated in the pathogenesis and progression of autoimmune diseases. Previous association studies have been performed to investigate the effect of polymorphisms in  Autoimmune diseases are a group of diseases characterized by abnormal immune response to autoantigens and self-tissue destruction , 2. It iα-amino acid that is an intermediary product in methionine metabolism MTHFR, lod widely , AS 29]MTHFR, lod widely . Previoud widely , and GD  MS risk . The conMTHFR polymorphisms with autoimmune diseases, including psoriasis, BD, GD, AS, and MS were identified before July 2021. The literature search was performed in the following electronic databases: PubMed, Ovid, Web of science, and China National Knowledge Infrastructure. The following items were used: “methylenetetrahydrofolate reductase”, “MTHFR”, “polymorphism”, “variant”, “SNP”, “psoriasis”, “Behcet's disease”, “Graves' disease”, “ankylosing spondylitis”, and “multiple sclerosis”. Additional articles were retrieved by checking references that were cited in the selected articles. No language restrictions were applied for the search strategy.The systematic review and meta-analysis followed the PRISMA 2020 statement guideline . ReportsMTHFR 677 C/T or 1298 A/C polymorphism with the following autoimmune diseases, including psoriasis, BD, GD, AS, and MS. We excluded studies that were case report, review articles, and records with insufficient data for calculating odds ratio (OR) and 95% confidence interval (CI). Studies that presented overlapping cases or controls were also excluded.We included case-control and case-cohort studies that investigated the association of the LM and PK reviewed and extracted simultaneously the data using a predefined form. The following information was extracted: the first author's name, year of publication, country of origin, ethnicity, sample size, definition of cases and controls, genotyping technique, and quality control of genotyping technique. Interreviewer disagreements were resolved by discussion with the third author (LY) who double checked the raw data.With reference to the Newcastle-Ottawa Scale, we assessed the quality of included studies by dividing them into two groups: high-quality  and low- quality (less than 4 stars) studies. The assessment was performed independently by LM and PK, and discrepancies were resolved by consultation with LY.Q-statistic test and I2 metric were used to assess the heterogeneity across studies [P value was more than 0.10 and I2 was less than 50%, indicating that there was no heterogeneity, the fixed-effect model was used; otherwise, the random-effect model was used [MTHFR polymorphisms and the risk of autoimmune diseases, four comparisons  were used by computing summary OR and 95% CI. Subgroup analyses based on racial decent (Asians and Caucasians) and study quality were also done. The robustness of results was evaluated using a sensitivity analysis when removing a single study at a time. Egger's linear regression asymmetry test was used to assess publication bias of included studies [P value < 0.05 was considered significant. All the statistical analyses were performed using STATA software version 11.0 .Hardy-Weinberg equilibrium (HWE) was assessed using a chi-squared test. The  studies . If the was used , 38. To  studies . A two-tZ-curve were constructed with a type I error of 5%, power of 80%, and relative risk reduction of 5-20%. Required information size (RIS) was also estimated. For the MTHFR 677 C/T polymorphism, TSA was performed under a recessive genetic model; for the 1298 A/C polymorphism, TSA was performed under a dominant genetic model.Trial sequential analysis (TSA) software version 0.9.5.10 beta was used to assess the reliability of the results from meta-analysis . The O'BMTHFR polymorphisms, lack of autoimmune diseases, absence of controls and available data, review articles, overlapping data, and case report. Finally, 43 studies met the inclusion criteria and were enrolled in the meta-analysis, including 14 investigating psoriasis, 10 investigating BD, 9 investigating MS, 5 investigating AS, and 3 investigating GD . The borderline statistical significance was also observed in Caucasians  rather than in Asians. In subgroup analysis based on study quality, no significant association of the MTHFR 677 C/T polymorphism with psoriasis risk was found in both high-quality and low-quality studies . In overall analysis, no significant association of the MTHFR 1298 A/C polymorphism with psoriasis risk was found. In sensitivity analysis, however, the exclusion of the study by Agha et al., Beranek et al., or Wu et al. yielded a significant result [Five studies investigated the association of the controls . Two stut result –46.MTHFR 677 C/T polymorphism with BD risk, involving 825 cases and 892 controls  and Caucasians . After subgroup analysis based on study quality, the significant association of the MTHFR 677 C/T polymorphism with BD risk was also found in low-quality studies  (MTHFR 677 C/T polymorphism with BD risk [Ten studies investigated the association of the controls . Among t03-2.54) . Sensiti BD risk .MTHFR 677 C/T polymorphism with MS risk, involving 1227 cases and 1426 controls  (Nine studies investigated the association of the controls . Meta-anctively) .MTHFR 1298 A/C polymorphism with MS risk, involving 952 cases and 1232 controls  and subgroup analysis, such as high-quality studies  and low-quality studies  (MTHFR 677 C/T and 1298 A/C polymorphisms increasing MS risk.Seven studies investigated the association of the controls . All thectively) . SensitiMTHFR 677 C/T polymorphism with AS risk, involving 545 cases and 502 controls . In subgroup analysis, the significant association of the MTHFR 677 C/T polymorphism with MS risk was also found in Asians , Caucasians , high-quality studies , and low-quality studies  (Five studies investigated the association of the controls . Among tctively) . SensitiMTHFR 677 C/T and 1298 A/C polymorphisms with GD risk   .Autoimmune disease is a pathological state that occurs when self- or autoantigens are misidentified as foreign entities by the immune system, resulting in tissue destruction and chronic inflammation , 2. AlthMTHFR deficiency is related to the development and progression of autoimmune diseases [MTHFR were considered to contribute to the occurrence of autoimmune diseases [MTHFR 677 T allele was associated with an increased risk of psoriasis [MTHFR 677 C/T with the etiology of psoriasis was found [MTHFR is a crucial enzyme in Hcy/folate metabolism, and the diseases , 17. Prediseases –33. The soriasis , whereassoriasis . Two preas found , 52. It as found , 45, 53.MTHFR 677 C/T has been examined extensively whether it influences individuals' susceptibility to BD and MS. However, conflicting results were also obtained. Karakus et al. and Naghibalhossaini et al. reported that subjects carrying the 677 T allele had an increased susceptibility to BD and MS [MTHFR 677 C/T was not a risk factor for BD and MS [MTHFR 677 TT genotype was associated with a higher risk of BD and MS. The discordant results may be due to small sample sizes in the meta-analysis reported by Chamorro et al. that included 494 BD patients and 374 controls [Apart from psoriasis, the D and MS , 47, wheD and MS , 54. TheD and MS , 55. Howcontrols , and theMTHFR 677 C/T and AS risk, 5 studies were included in this meta-analysis. The pooled analysis showed a significant association between AS and the MTHFR 677 TT genotype in overall comparison. Subgroup analyses based on ethnicity and study quality also showed the significant association. Notably, carriers with the MTHFR 677 TT genotype had a 2.48-increased risk of AS risk in high-quality studies under a recessive genetic model. To the best of our knowledge, this is the first meta-analysis to assess the effect of the MTHFR 677 C/T polymorphism on AS risk, which provides stronger evidence than any previous case-control study.Regarding the association between the MTHFR 677 C/T and GD risk. When we pooled all data together, only 409 cases and 418 controls were included in the current meta-analysis. TSA also showed that the findings may be not conclusive, and thus, the negative results in the current study should be interpreted with caution. Additional studies are needed to be performed to clarify the exact role of the MTHFR 677 C/T in the pathogenesis of GD.To date, only 3 studies investigated the association between the MTHFR 677 C/T, the 1298 A/C polymorphism was reported to be associated significantly with MS risk [MTHFR 1298 A/C polymorphism and MS was found [MTHFR 1298 AC and AC/CC genotypes were associated with increased risks of MS both in overall comparison and subgroup analysis according to study quality. Sample size is an important issue for a sound association study because small samples may result in insufficient power to obtain the real effect. Our study has more samples than previous meta-analysis reported by Lee et al. [Besides the  MS risk –58. Howe MS risk . A meta-as found . In the as found , 56, 59 e et al. , indicatMTHFR 1298 A/C polymorphism was also analyzed in psoriasis and GD patients. Not surprisingly, inconsistent results were also observed. Kilic et al. reported that the prevalence of the 1298 C allele was higher by 17.0-fold in patients with psoriasis compared to the control group [MTHFR 1298 A/C polymorphism and GD risk, only 3 studies were included in this meta-analysis and no significant association was observed. According to the data from TSA, however, we cannot conclude that the 1298 A/C was not a risk factor for GD since the limited samples may lead to false negative error.The ol group , whereasol group . In thisIn this meta-analysis, heterogeneity across studies was observed in some comparisons. Unfortunately, we did not identify what caused the heterogeneity. Despite the disadvantages, no publication bias is present, suggesting there is no evidence of publication selection bias in the literature.MTHFR 677 C/T with psoriasis was borderline significant. After the exclusion of the study by Izmirli et al., Luo et al., or Wang et al., the findings were changed [MTHFR 677 C/T and 1298 A/C polymorphisms and GD risk was investigated only in a few studies [MTHFR on the risk of autoimmune diseases.This study contains some limitations that should be discussed. The association of the  changed –43, sugg studies , 60, 61;MTHFR 677 C/T was a risk factor for BD, MS, and AS and the 1298 A/C was a risk factor for MS. Considering the small number of studies included in some certain subgroups, further larger-sample investigations performed in diverse ethnic groups are necessary to confirm these findings. Functional analyses are also required to elucidate whether and how the SNPs in MTHFR impact the etiology of autoimmune diseases.In conclusion, this meta-analysis demonstrated that the


Endometriosis (EMS) is confirmed pathophysiologically to be an estrogen-dependent disease, similar to endometrial hyperplasia/cancer and breast cancer. Epidemiological and biological data on endometriosis might explain links between endometriosis and these cancers. We sought to identify the differences in the risk of endometrial cancer and breast cancer between women with and women without endometriosis.2 were used to evaluate the degree of heterogeneity of eligible studies. We used funnel plots and Begg’s and Egger’s tests to assess publication bias.We searched PubMed, EMBASE, the Cochrane Library, and four Chinese databases  to identify relevant studies published online between January 2011 and March 2021. In our meta-analysis, we used the Newcastle–Ottawa Scale (NOS) to evaluate the design and quality of all studies, and we calculated the pooled risk ratio (RR) using the random model. The Q test and IOf the 1369 articles, we finally included 14 cohort studies and seven case–control studies. Data from large cohort and case–control studies indicate that women with endometriosis had an increased risk of both endometrial cancer  and breast cancer .Endometriosis can increase the risk of endometrial cancer and breast cancer, and women with endometriosis are recommended to receive routine screening in long-term management.The online version contains supplementary material available at 10.1186/s12905-022-02028-x. Endometriosis (EMS) is a common inflammatory condition defined as endometrial-like tissues found outside of the uterus, mainly in the pelvic area . Three well-recognized subtypes are named superficial endometriosis (SUP), ovarian endometrioma (OMA), and deep infiltrating endometriosis (DIE) . Pelvic Although endometriosis is a benign gynecological disease, the pattern of its growth is similar to that of malignant disease . The ectThe retrograde menstruation hypothesis, which is commonly accepted, posits that the mechanism of endometriosis is that eutopic endometrial tissues with molecular defects migrate retrogradely to the abdominal cavity mixed with blood, stick to the peritoneum, and proliferate aggressively, and that endometriosis  can also be regulated by hormones. Therefore, the association between endometriosis and endometrial cancer seems to be noticed by researchers more easily, while breast cancer, which is the other common cancer among reproductive women, can also be influenced by hormone fluctuations. Endometriosis itself and before or after therapy may influence breast cancer directly or indirectly. Epidemiologically, several studies have clearly shown that endometriosis is a risk factor for ovarian cancer , 17, 18,The reporting of this meta-analysis strictly followed the MOOSE checklist. We comprehensively searched for published relevant observational studies from the databases of PubMed, EMBASE, Cochrane Library, CNKI, VIP, WanFang, and CBM for the past 10 years . The search terms were the keywords combined with their corresponding MeSH terms (which are detailed in the We included relevant studies that met the following criteria: (1) studies that examined endometriosis  and endometrial cancer or breast cancer; (2) human studies and cohort or case–control studies; and (3) publications in which usable risk estimates, such as odds ratios (ORs), risk ratios (RRs), hazard ratios (HRs), and standard incidence ratios (SIRs) with 95% confidence intervals (CIs), were presented or necessary data were given for calculation. (4) If several studies were conducted in the same population, we would select the report with the most applicable estimates or the most recent report. However, we also excluded the following types of studies: (1) meta-analyses, reviews, case reports, editorials, and letters to the editor; (2) animal or cell experiments; (3) studies not published in English or Chinese; and (4) studies not meeting the inclusion criteria.Based on the predetermined selection and exclusion criteria, two authors independently extracted data from each study using predesigned forms, and discrepancies were resolved by the third author. For each study, we independently extracted the first author’s name, year of publication, country, age at baseline of the study population, study design, follow-up time of the cohort studies, the sample size of the study, ascertainment of endometriosis and cancer cases, age at the diagnosis of endometriosis, adjustment factors and relative risk estimates with 95% CIs (we chose the model adjusted for the largest number of confounders when different crude and adjusted estimates were reported), and the method of information collection.The Newcastle–Ottawa Scale (NOS) was used to assess the quality of the included cohort and case–control studies . The NOSQ test and I2 were used to evaluate the degree of heterogeneity of eligible studies. For the Q test, p > 0.10 was considered representative without statistical heterogeneity, data were interpreted using the fixed effect model, p < 0.10 indicated statistically significant heterogeneity, and the random effects model was chosen. For I2, the values of 0, 25, 50, and 75% correspond to no, low, moderate, and high heterogeneity, respectively   and p = 0.629 (> 0.05), respectively, suggesting that there was no publication bias of the included studies.Publication bias was assessed by the funnel plot, as shown in Fig. We used leave-one-out sensitivity analysis to evaluate whether any small study effect influenced the pooled effect size. As a result, no significant changes were observed in the sensitivity analysis Fig. , suggestQ = 86.62, P = 0.000; I2 = 82.7%). In the subgroup analysis, the group stratified by the ascertainment of breast cancer, ascertainment based on histopathology reports  and other methods  showed low heterogeneity. Studies adjusted for oral contraceptive use and pregnancies also showed low histopathology, which may have been the source of heterogeneity ]  and p = 0.698 (> 0.05), respectively, suggesting that there was no publication bias. In the leave-one-out sensitivity analysis, no significant changes were observed  classification system for EMS, proposed in 1979 and revised in 1985, is commonly used in the clinic. According to its total scores, EMS is classified into 4 stages: 1–5 denotes stage I ; 6–15 denotes stage II (mild lesions); 16–40 denotes stage III (moderate lesions); and 41–150 denotes stage IV (severe endometriosis) , 48. A sLike endometrial cancer and breast cancer, endometriosis is also an estrogen-dependent disease. In endometriotic tissue, aromatase stimulated by PGE2 (Prostaglandin E2), which is essential in the compounding of estrogen, is present at high levels, as well as a lack of 17β-HSD (17β-hydroxysteroid-dehydrogenase) type 2, which can convert estradiol (E2) to the less potent estrone (E1), leading to the accumulation of excess estrogen . In addiFor endometriosis populations, the combined oral contraceptive pill (COCP) is the first-line drug to relieve symptoms and is widely used in adolescents < 16 years of age. COCP is a compound preparation of a certain amount of estrogen and progesterone that can directly act on the endometrium and can simultaneously act on the hypothalamus through negative feedback and inhibit the secretion of gonadotropin-releasing hormone (GnRH). Studies show that COCP can decrease the risk of endometrial cancer, which means that for people with endometriosis, it may protect the endometrium from malignant transformation , 61, butKnowing the association between endometriosis and cancer has important public and prevalent clinical implications. Our meta-analysis clearly showed that endometriosis can increase the risk of endometrial cancer and breast cancer, which may be significant for long-term management, but we cannot ignore the between-study heterogeneity, which may influence the credibility of the results of our study. For future research, we should perform further stratification research based on the AFS stage or macrophenotype, restricting the increased risk of cancer to specific populations, which may be more valuable for regular screening.Additional file 1. Supplementary files NO.1 Basic characteristics and quality evaluation.Additional file 2. Supplementary files NO.2 Search strategy.Additional file 3. Supplementary files NO.3 Stata analysis.


AR) have been reported to contribute to the pathogenesis of epilepsy and the recurrence of chronic seizures. Genetic polymorphisms in GABRA1 and GABRA6 may confer a high risk of epilepsy and multiple drug resistance, but with conflicting results. We aimed to assess the association of GABRA1 rs2279020 and GABRA6 rs3219151 with epilepsy risk using a meta-analysis. The databases of Pubmed, Ovid, Web of Science, and China National Knowledge Infrastructure were searched. Summary odds ratios (ORs) and 95% confidence intervals (CIs) were computed to evaluate the association between the polymorphisms and epilepsy risk using a fixed- or random-effect model. Trial sequential analysis (TSA) was performed to assess the results of the meta-analysis. No significant association between the GABRA1 rs2279020 and GABRA6 rs3219151 and the risk of epilepsy was found in the Asian and Arabic populations. The negative results were also observed when comparing the GABRA1 rs2279020 and GABRA6 rs3219151 polymorphism to antiepileptic drug responsiveness. The trial sequential analysis confirmed the results of the meta-analysis. This meta-analysis suggests that GABRA1 rs2279020 and GABRA6 rs3219151 are not risk factors for the etiology of epilepsy and antiepileptic drug responsiveness in the Asian and Arabic populations.The γ-aminobutyric acid type A receptors (GABA Among all neurological diseases, epilepsy is the most common one with over 50 million people diagnosed worldwide , 2. It iAR) play pivotal roles in the suppression of epileptogenesis by maintaining homeostasis over brain excitation (AR subtypes may be involved in drug resistance (Antiepileptic drugs (AEDs) were commonly used to prevent epileptic seizures. Almost one-third of patients, however, continue to show recurrent seizures despite optimal AEDs treatment . Geneticcitation –10. In asistance , 10.ARs are heteropentamers encoded by 19 subunits of α (1–6), β (1–3), γ (1–3), δ, ε, θ, π, and ρ (1–3)  . The α1β ρ (1–3) , 8. In hseizures . Single seizures –15. For sistance , whereassistance . Given tGABRA1 or gamma-aminobutyric acid type A receptor subunit alpha6 or GABRA6) and (polymorphis* or variant* or SNP*) and (epilepsy). The literature screening was performed until 18 May 2022 without language and ethnic restrictions. If the language of publication was non-English, we translated the required information into English.The methodology of this meta-analysis followed the PRISMA 2020 statement guidelines. We identified records from the databases of Pubmed, Ovid, Web of Science, and the China National Knowledge Infrastructure using the following searching strategies:  contained sufficient original data to compute summary odds ratios (ORs) and 95% confidence intervals (CIs). The exclusion criteria were as follows: (a) review articles; (b) absence of available data; and (c) overlapping data reported by the same research group.The inclusion criteria were as follows: (a) investigated the association of GABRA1 rs2279020 and GABRA6 rs3219151 polymorphisms.ZT and YY independently extracted the following data from the included studies: the name of the first author, year of publication, ethnicity of study population, number of cases and controls, criteria used to define drug-responsiveness and drug-resistance, and genotyping technique and genotype distributions of ZT and YY independently assessed the eligible studies using the modified Newcastle-Ottawa Scale . StudiesGABRA1 rs2279020 and GABRA6 rs3219151 polymorphisms with the risk of epilepsy and drug resistance, TSA software (version 0.9.5.10 beta) was performed to evaluate the reliability of the meta-analysis under a recessive model. The O'Brien-Fleming boundary, sequential monitoring boundary, futility boundary, and cumulative test statistic (Z-curve) were calculated with an alpha of 5%, power level of 80%, and relative risk reduction of 20% , the Hardy–Weinberg equilibrium (HWE), and study quality (high and low). Cochran's Q-test and I2 statistics were used to assess the inter-study heterogeneity (Q statistic (P < 0.10) indicated high heterogeneity. Therefore, a random-effects model was used for the meta-analysis . The pooled odds ratio (OR) and its 95% confidence interval (CI) were calculated to assess the association of the GABRA1 rs2279020 and GABRA6 rs3219151 polymorphisms (n = 21), lack of available data (n = 8), review articles (n = 8), and overlapping data (n = 1). Finally, 13 studies were included in this meta-analysis, including 11 investigating GABRA1 rs2279020 polymorphism and five investigating GABRA6 rs3219151 polymorphism  of the studies used the second genotyping method for quality control in this study.The characteristics of the included studies are summarized in GABRA1 rs2279020 polymorphism, eight studies were included involving 1,688 patients with epilepsy and 1,233 controls. Among them, 75% (6/8) were performed in Asians and 25% (2/8) were performed in Arabians; 62.5% (5/8) were high-quality studies and did not derive from HWE in controls. In an overall analysis, no significant association between the GABRA1 rs2279020 and the risk of epilepsy was found. Similarly, the negative results were also observed in subgroup analyses based on ethnicity and study quality. Even though the GABRA1 rs2279020 was associated with an increased risk of epilepsy among studies without HWE under heterozygous comparison  and dominant genetic model , the polymorphism was not associated with the risk of epilepsy among studies with HWE in controls  but not in the high-quality studies  (For the 01–1.72) .GABRA1 rs2279020 polymorphism with epilepsy risk was observed when the study by Balan et al. was excluded from the heterozygous comparison  was reported to be potentially functional, with the A → G shift changing the shape of the protein that led to alternative splicing , the Key Project for Science and Technology Department of Sichuan Province (Grant No. 2018SZ0216), and the Sichuan University (Grant No. 2017scu11038).The authors declare that the research was conducted in the absence of any commercial or financial relationships that could be construed as a potential conflict of interest.All claims expressed in this article are solely those of the authors and do not necessarily represent those of their affiliated organizations, or those of the publisher, the editors and the reviewers. 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