UNLABELLED: Microbial communities have an important role in natural ecosystems and have an impact on animal and human health. Intuitive graphic and analytical tools that can facilitate the study of these communities are in short supply. This article introduces Microbial Community Analysis GUI, a graphical user interface (GUI) for the R-programming language (R Development Core Team, 2010). With this application, researchers can input aligned and clustered sequence data to create custom abundance tables and perform analyses specific to their needs. This GUI provides a flexible modular platform, expandable to include other statistical tools for microbial community analysis in the future. AVAILABILITY: The mcaGUI package and source are freely available as part of Bionconductor at http://www.bioconductor.org/packages/release/bioc/html/mcaGUI.html

Cho and Lee (2004) proposed a Bayesian hierarchical error model (HEM) to account for heterogeneous error variability in oligonucleotide microarray experiments. They estimated the parameters of their model using Markov Chain Monte Carlo (MCMC) and proposed an F-like summary statistic to identify differentially expressed genes under multiple conditions. Their HEM is one of the emerging Bayesian hierarchical modeling tools that have been developed for the analysis of multiple-level data structures and variation in microarray gene expression data (Broet et al., 2002; Tadesse and Ibrahim, 2004; Cho and Lee, 2004). In this letter, we first discuss the significance of the HEM developed by Cho and Lee. Then, we re-derive the fully conditional distributions for gene and conditional effects, since we think that these two fully conditional distributions were not presented properly in their paper. Finally, we expand the HEM to deal with biological or/and experimental correlations in gene expression data. A FORTRAN 90 program was developed to implement our extended method and it is available from the authors upon request. Significance of HEM. Many aspects of the HEM presented in the article of Cho and Lee (2004) are significant. The following are the two aspects that we believe are important in the analysis of microarray gene expression data. First, error variability of gene expression data is decomposed into two components, biological and experimental. The former reflects variation owing to biological sample heterogeneity that affects gene expression, while the latter largely comes from variation in sample preparation and sample analysis. Estimating these two sources of errors separately will not only improve estimation of model parameters but also facilitate a better understanding of patterns of over all gene expression. Comparison of Markov chain sampling of a gene effect using (a) previous algorithm by Cho and Lee (2004) and (b) corrected algorithm by Wu, Forney and Joyce. The simulated gene effect was 0.7923. The y-axis represents posterior estimates of the gene effect whereas the x-axis represents saved (i.e. every one-tenth) Markov chain sampling steps. Modeling biological/experimental correlations. Cho and Lee (2004) assumed independence for both biological and experimental errors, but these assumptions may be relaxed in the HEM. In this section, we expand their HEM to model biological or/and experimental correlation in gene expression data. First, consider correlations among experimental replicates, which may arise when experimental replicates are exposed to some common environmental factors yet collected at different times or locations. In these cases, the gene expression data may show temporal or spatial variation. In bacterial biofilms, for example, microorganisms undergo significant phenotypic changes during the transition from planktonic to biofilm forms, which reflects altered underlying gene expression and regulation as they adapt to form highly organized and structured sessile communities (Schembri et al., 2003; Watnick and Kolter, 2000) with enhanced resistance to antimicrobial treatments and host immune responses. Furthermore, the development of a biofilm may be tuned differently as a response to a nutrient gradient from the bulk media to the inside of the biofilm. Thus, gene expression at different locations in a biofilm may show spatial variation similar to geographic data in such a way that ‘everything is related to everything else, but near things are more related than distant things’ (Tobler, 1970). Biofilms are the dominant mode of growth for bacteria in the environment, and are of great concern in medical, environmental and industry settings. Therefore, considering the spatial correlation in a model would help better understanding the formation of biofims as well as their altered biological activities. Our extended model facilitates estimating patterns of biological or/and experimental (co)variance or correlations (Fig. 2) whereas such information could not be inferred using the previous model (Cho and Lee, 2004). As mentioned before, biological or experimental correlations could be of considerable interest and necessary to help us better understand the nature of correlated gene expression. For example, suppose the pattern of experimental (co)variance in Figure 2b represents the environmental correlations in a biofilm. We might conclude that nutrient differences result in higher correlations between the inner and bottom layers than between the top and the other two layers. 3D patterns of estimated (co)variance components among (a) five biological sample and (b) three experimental replicates. Further, we investigated how different model specifications about biological/experimental covariance (i.e. zero versus non-zero) would affect the posterior estimates of parameters of interest. Ignoring experimental covariance in the model did not lead to significant differences in parameter estimation (data not shown). However biological covariance components are important for estimating gene and interaction effects. If the biological covariance components are ignored then gene effects and interaction effects will be estimated with large biases. However, grand mean, condition effects and the common residual variance appear to be more robust, and gave similar results under either model specification about biological (co)variance. As shown in Table 1, considering non-zero biological (co)variance in the model considerably decreased estimation biases of gene and interaction effects. Squared biases, and consequently MSE, of these parameters were smaller in the analysis considering biological covariance components than by ignoring them. Given the variation that occurs during the collection of gene expression data, posterior estimation of gene and interaction effects was better, in terms of MSE, sampling variation and squared biases, when more samples were used (Table 1). This result would challenge the present practice where a very limited number (say three) of samples are used in most gene expression DNA array experiments. Mean squared error (MSE), sampling variance (SV) and squared bias (SB) in posterior estimation of model effects under the assumption of zero [COV(b) = 0] versus non-zero [COV(b) ≠ 0] biological covariancea,b aμ = grand mean; gi = effect of gene i; cj = effect of condition j; rij = interaction between gene i and condition j; V(b)= variance–covariance matrix of biological errors; σe2 = variance of residual errors. bB05R03 = five biological samples each with three experimental replicates; B10R06 = 10 biological samples each with 6 experimental replicates. Mean squared error (MSE), sampling variance (SV) and squared bias (SB) in posterior estimation of model effects under the assumption of zero [COV(b) = 0] versus non-zero [COV(b) ≠ 0] biological covariancea,b aμ = grand mean; gi = effect of gene i; cj = effect of condition j; rij = interaction between gene i and condition j; V(b)= variance–covariance matrix of biological errors; σe2 = variance of residual errors. bB05R03 = five biological samples each with three experimental replicates; B10R06 = 10 biological samples each with 6 experimental replicates. Finally, we show how model specifications about biological covariance would affect the estimation of gene expression difference, (μ + gi + cj + rij) – (μ + gi + cj′ + rij′) = (cj–cj') + (rij–rij') between various conditions (Fig. 3), which is of importance in identifying differential gene expression. Clearly, the estimated gene expression differences were in better agreement with true differences when considering non-zero biological covariance as compared to ignoring them in the model. In the former case, the R-square value for the linear regression of the estimated and true expression differences was higher and the regression slope is closer to 1.0. The low slope in Figure 3b suggests an obvious systematic bias, which resulted from ignoring non-zero biological covariance. On average, the higher the biological correlations were, the greater the difference. In 10 independent simulations starting with different datasets, we found that 6.5–38.7% of the top 100 differentially expressed genes could vary when one assumes zero or non-biological covariance, respectively. Regression of estimated to true gene expression difference obtained by assuming (a) zero versus (b) non-zero biological covariance components. In conclusion, we extended the model of Cho and Lee (2004) to allow the analysis of gene expression data with biological/environmental correlations, which not only facilitate estimating biological/expression correlation patterns, but it also improved estimation of model parameters and consequently identifying differentially expressed genes in DNA array experiments. A FORTRAN 90 program was developed to implement our extended method and it is freely downloadable at Author Webpage