- Methodology article
- Open Access
Genome sequence-based species delimitation with confidence intervals and improved distance functions
- Jan P Meier-Kolthoff^{1},
- Alexander F Auch^{2},
- Hans-Peter Klenk^{1} and
- Markus Göker^{1}Email author
https://doi.org/10.1186/1471-2105-14-60
© Meier-Kolthoff et al.; licensee BioMed Central Ltd. 2013
- Received: 26 November 2012
- Accepted: 4 February 2013
- Published: 21 February 2013
Abstract
Background
For the last 25 years species delimitation in prokaryotes (Archaea and Bacteria) was to a large extent based on DNA-DNA hybridization (DDH), a tedious lab procedure designed in the early 1970s that served its purpose astonishingly well in the absence of deciphered genome sequences. With the rapid progress in genome sequencing time has come to directly use the now available and easy to generate genome sequences for delimitation of species. GBDP (Genome Blast Distance Phylogeny) infers genome-to-genome distances between pairs of entirely or partially sequenced genomes, a digital, highly reliable estimator for the relatedness of genomes. Its application as an in-silico replacement for DDH was recently introduced. The main challenge in the implementation of such an application is to produce digital DDH values that must mimic the wet-lab DDH values as close as possible to ensure consistency in the Prokaryotic species concept.
Results
Correlation and regression analyses were used to determine the best-performing methods and the most influential parameters. GBDP was further enriched with a set of new features such as confidence intervals for intergenomic distances obtained via resampling or via the statistical models for DDH prediction and an additional family of distance functions. As in previous analyses, GBDP obtained the highest agreement with wet-lab DDH among all tested methods, but improved models led to a further increase in the accuracy of DDH prediction. Confidence intervals yielded stable results when inferred from the statistical models, whereas those obtained via resampling showed marked differences between the underlying distance functions.
Conclusions
Despite the high accuracy of GBDP-based DDH prediction, inferences from limited empirical data are always associated with a certain degree of uncertainty. It is thus crucial to enrich in-silico DDH replacements with confidence-interval estimation, enabling the user to statistically evaluate the outcomes. Such methodological advancements, easily accessible through the web service at http://ggdc.dsmz.de, are crucial steps towards a consistent and truly genome sequence-based classification of microorganisms.
Keywords
- Archaea
- Bacteria
- BLAST
- DDH
- GGD
- GGDC
- GBDP
- Genomics
- MUMmer
- Phylogeny
- Species concept
- Taxonomy
Background
DNA-DNA hybridization (DDH) is a wet-lab method currently still used as the taxonomic gold standard for species delineation in Archaea and Bacteria. If the genomic DNA of two respective organisms reveals a DDH similarity of below 70% this is the main argument to regard them as distinct species and vice versa [1, 2]. DDH is widely considered as tedious, laborious and potentially rather error-prone [3, 4]. Moreover, in contrast to genome sequencing it does not return more information than the DDH value itself and, as a consequence, it is impossible to work incrementally by re-using data.
The DDH technique is currently established in only a few specialized labs (mainly microbial service collections) and, because it is prone to experimental deviation, requires several experimental repetitions to determine the statistical confidence of that experiment. For instance, regarding species delimitation in microbiology, the relevant question is whether or not the DDH value is significantly below or above 70%. This is particularly important in the context of a polyphasic approach, in which the evidence from DDH has to be traded off against other criteria such as phenotypic measurements [5]. DDH experiments can be omitted in descriptions of novel species only if the 16S rRNA sequence similarity is below a certain threshold, indicating that DDH values above 70% cannot be expected [2].
The increasing availability of genome sequences thus triggered the development of computational techniques to replace wet-lab DDH [6-8]. These were expected to provide the deepest possible resolution for differentiation, to ensure much higher reproducibility of the results and to allow incremental work by filling databases with type-strain genome sequences [4]. But unless high correlations with wet-lab DDH, and precise models for estimating DDH or at least DDH-analogous species boundaries from genome-to-genome comparisons, were available, the newly calculated values were not comparable to the previous ones and could yield largely deviating species-boundary estimates and, thus, an inconsistent microbial taxonomic classification. Hence, for obvious reasons the literature on in-silico replacements for DDH considered correspondence with wet-lab DDH values as optimality criterion. As a consequence, regression and/or correlation analyses with wet-lab DDH values were used throughout for the calibration and optimization of the in-silico replacement methods [6-8].
In view of the technical problems and progress the relation between the wet-lab DDH procedure and digital estimation of DDH equivalents reminds very much to what happened some 30 years ago when DNA:rRNA cross-hybridization melting curves [9, 10] were replaced by 16S rRNA sequences, which supported a significant progress in microbial phylogeny [11].
The Genome Blast Distance Phylogeny approach (GBDP) was originally devised as an approach for the inference of phylogenetic trees or networks from a given set of wholly (or even incompletely) sequenced genomes [12], and was subsequently revisited and enhanced [8, 13-16]. The underlying principle is as follows: in the first step two genomes A and B are locally aligned using tools such as BLAST[17], which produce a set of high-scoring segment pairs (HSPs; these are intergenomic matches). In the second step, information contained in these HSPs (e.g., the total number of identical base pairs) is transformed into a single genome-to-genome distance value by the use of a specific distance formula. Phylogenetic trees can then be inferred from such distance matrices using standard techniques such as neighbour joining [18]. These methods are robust even in the presence of a significant amount of paralogous genes, large repeats and reduced genomes [12], as well as low complexity-regions within the sequences [16]. GBDP could also be applied to proteomic data [13] and even to single genes [19].
A further use of GBDP was recently evaluated, namely to infer digital equivalents for DDH values [8, 16]. These turned out to successfully mimic the wet-lab hybridization results, providing higher correlations with an empirical set of DDH values than antecedent genome sequence-based methods [6] and being able to deal with rather incomplete genomes [8]. Microbiologists can make use of GBDP by means of a free web service at http://ggdc.dsmz.de[16] for submitting genome pairs and receiving DDH analogues as well as model-based DDH estimates. Such values on the original scale of wet-lab DDH measurements have the practical advantage that the well-known 70% threshold can still be applied [8], even though they are mathematically equivalent to in-silico DDH analogues that use a scale of their own, and accordingly represent a novel species-delimitation threshold [6, 7].
The first goal of the present study is to improve DDH estimation from genome-sequence comparisons by using a more comprehensive empirical database and by considering a broader range of numerical data transformations and statistical models. Previous studies were limited to regression models of the untransformed data and thus presupposed a linear relationship between wet-lab DDH and the results of genome-sequence comparisons [6-8]. But this assumption might be unjustified, and the inspection of more complex models ([20], pp. 58-79) and distinct data transformations [21, 22] has frequently been recommended. In addition to suboptimal fits, linear models can lead to DDH predictions below 0% and above 100% similarity if the underlying (dis-)similarity values are close to the upper bound (1 if distances are not logarithmically transformed) or the lower bound (0), respectively.
The second goal of these examinations is to obtain confidence intervals for in-silico DDH values - an indicator showing taxonomists how uncertain a reported value is, especially if it is close to the 70% boundary. Even though it is safe to assume a priori that digital DDH values display much less variability than wet-lab DDH experiments given the high sequence coverage that can be obtained with state-of-the-art sequencing technology [4], it is of interest whether, and how, confidence intervals can be calculated for the in-silico replacement methods, too. Hence, GBDP was further extended by integrating resampling techniques for calculating a confidence interval per point estimate (i.e., pairwise distance). Bootstrapping [23] and jackknifing [24] are well-known and robust resampling techniques for estimating the variance of a sample. But uncertainty might additionally or mainly be caused by the empirical modeling of the relationship between DDH and genome-sequence comparisons (see below), and the relative proportions of resampling- and model-based confidence intervals need to be assessed. Confidence intervals would render GBDP the first in-silico procedure to infer DDH analogs that can be statistically evaluated, which is particularly important in the context of the polyphasic approach to microbial taxonomy (see above).
The third topic of this study is to broaden the range of considered GBDP distance functions. Whereas [8] already investigated a much more diverse range of in-silico DDH analogues than previous publications on DDH-replacement methods [6, 7], here GBDP is further enriched with so-called “coverage distances” [12] (and bootstrapping and jackknifing for this algorithm). The performance of the novel GBDP implementation could thus be assessed under 4350 distinct settings (ranging from the local alignment tools and their settings to the distance functions), requiring a total of 136 million individual genome comparisons (including bootstrapping and jackknifing), and the overall best-performing settings determined. The effects of the parameters used for the calculation of intergenomic distances on the resulting correspondence with the DDH values was also investigated in detail using multiple regression.
The results of this study are thus likely to contribute toward progress in using the comprehensive information encoded in entire genomes for the taxonomy of prokaryotes.
Methods
Extended benchmark data set
The DDH benchmark data set was extended compared to previous studies aiming at an increased precision and significance of the ranking of the genome-to-genome distance methods and the models for the conversion to DDH values. In detail, the here used data set (henceforth called “DS1”) comprised 156 unique genome pairs along with their respective DDH values: 62 from Goris et al. [6], 31 from the GOLD database [25], and 63 from Richter et al. [7]. Only the first two sources had been considered in a previous publication on GBDP as DDH replacement [8].
If several DDH/ANIb/ANIm/Tetra values were present for a single genome pair, they were averaged. A single genome pair showed a DDH value above 100% similarity (i.e., 100.9% between Escherichia coli O157:H7 EDL933 and Escherichia coli O157:H7 Sakai). As it biologically made not much sense this value was set to 100% to maintain proper input data for some of the statistical models (see below). Another genome pair (Thermotoga maritima MSB8 and Thermotoga petrophila RKU-1) had a contradicting relation between its DDH value (16.9%) and the genome based distance/similarity measures (GBDP, ANI, ANIb, ANIm and Tetra) on the other hand [7]. Following [7], this questionable data point was excluded from the correlation analyses. The full list of genome pairs used in this study is found in the Additional file 1.
To detect significant deviations, if any, between the new and the previous GBDP implementation, the data subset “DS2” was created, containing only the previously available data points [8]. For comparing GBDP with the first ANI implementation, data subset “DS3” comprised the 62 data points in common between [6, 8]; for comparison with the JSpecies study, subset “DS4” contained only the 98 DDH values in common between [7, 8].
The GBDP principle, and its technical update
To motivate the upcoming changes such as the addition of support for BLAST+[26] and the completion of the implementation of the so-called “coverage” algorithm [12], the major steps within the GBDP pipeline [8, 12, 13] are summarized in the following.
The pipeline is primarily subdivided into two phases. First, a genome X is BLAST ed against a genome Y and vice versa (here, the term “BLAST ed” denotes the application of one out of six supported local-alignment programs; a full list of these programs is found in Additional file 2). BLAST+ has recently been added to the list of available programs, because it provides substantial speed improvements for long queries and database sequences [26]. The alignment process is done in one pass using all the available sequence information of both respective genomes, i.e., GBDP does not require the sequences to be artificially cut into pieces as do other approaches [6].
The resulting matches between both genomes are called high-scoring segment pairs (HSPs) and represent local alignments that are considered statistically significant if the associated expect value (e-value) is sufficiently low [27] (reliable thresholds are usually equal or less than 10^{−2}, but GBDP conducts the filtering itself, and its effect is explicitly addressed below).
The web service at http://ggdc.dsmz.de makes use of these formulae; the other GBDP formulae are minor variants [8, 12, 13] and are found in Additional file 3:
Each was devised to consider distinct aspects of intergenomic relationships. Formula d_{6} preserves most information, because it is some kind of combination of d_{0} and d_{4}. It also performs best in a phylogenetic context [13]. However, d_{4} is immune against problems caused by incompletely sequenced genomes, as it does not consider the genome lengths [8]. It follows from d_{0} to d_{6} that similarity instead of dissimilarity (distance) values could be easily obtained by subtracting the distances from 1 ([28], pp. 252-259) [8, 13]. This would be mathematically analogous for the subsequent correlation analyses because only the sign changed, but using distances is more convenient for inferring GBDP trees [12, 13].
Analogously, the number of genome positions covered by HSPs, H_{ X Y }and H_{ Y X }, as used in formulae 6 and 8, is calculated by counting all non-zero positions in the vectors v_{ X }and v_{ Y }, respectively. The coverage method is faster than the other two approaches, because there is no overhead caused by HSP sorting and/or trimming algorithms.
Conducting genome comparisons for the correlation analysis
A correlation analysis was conducted to show the overall performance of the GBDP method and to yield the best GBDP parameter setup. Six local-alignment tools were tested for genome comparisons: BLAST+[26], NCBI-BLAST[17], MUMmer[29], BLAT[30], WU-BLAST[17] and BLASTZ[31]. These were not only used to conduct six genome comparisons per available genome pair but were also applied in several passes, each time changing a chosen parameter, presumably affecting the local alignment and thus potentially improving the correlation with DDH. A special focus was on finding influential parameters for BLAST+ because it is one of the technically most advanced local-alignment tools available [26]. Moreover, all available distance functions and HSP filtering approaches were applied to each genome comparison (see above). Thus, all in all 4350 distinct settings were investigated, i.e., the product of 145 alignment settings, three algorithms for dealing with overlapping HSPs, and ten distance formulae. This resulted in a total of 136 million GBDP-based genome comparisons which had to be conducted. Additional file 2 provides a complete overview on these settings and numbers.
In general, studies of that kind are computationally challenging, because a huge number of input and result files need to be processed. This gave rise for equipping the method with an extension allowing it to be executed on compute clusters [32-34].
Analyzing correlations between intergenomic distances and DDH values
According to [8] Pearson’s ρ ([35], pp. 533) and Kendall’s τ ([28], pp. 198-199) were computed for all distinct GBDP settings and the DDH values, respectively. The necessary analysis pipeline was implemented as an R[36] script and applied to the previously described main data set DS1 and its subsets DS2-4. As GBDP-derived values are distance measures, whereas ANI values are similarities between genomes, the correlation coefficients’ sign had to be inverted accordingly to allow for the direct comparison of the performance of GBDP, ANI and JSpecies. The full list of the genome pairs and their associated DDH and ANI values is found in Additional file 1.
For GBDP, the influence of four predictor variables was tracked by means of multiple linear regression ([37], pp. 387-448): alignment method, algorithm for treating overlapping HSPs, distance formula and e-value filter method (Additional file 2). The identification of the least complex model that still explained most of the variation in the data was conducted with the R package MASS[38] under both forward variable selection and backward elimination. For an interpretation of possible interaction effects, the effects package [39] for R was used. Additionally, each predictor variable’s relative importance index [40] was computed.
GBDP bootstrapping and jackknifing
To obtain confidence-interval (CI) estimates, GBDP was augmented with sampling with replacement (bootstrapping [23]) and sampling without replacement (jackknifing [24]), by default using 50% deletion in each replicate. For the variants of GBDP that use a filtering approach for removing overlapping (parts of) HSPs, namely the “greedy” and “greedy-with-trimming” algorithms (see above), the implementation treats individual HSPs as the observations for resampling. After applying the filtering step, the original distance (the point estimate) is inferred from the set of filtered HSPs using one of the distance formulae described above. Then the set of HSPs is bootstrapped or jackknifed, and in each replicate a distance value is calculated using the same formula. For the “coverage” algorithm (see above), resampling was implemented by bootstrapping (or jackknifing) the constructed coverage vectors. For the evaluation of the methods, 100 replicates were used throughout.
The dependency of the resulting bootstrapping and jackknifing CIs on each genome pair’s original distance (point estimate) was investigated, as well as the effect of the GBDP method on the replicates’ variation. The latter was assessed via the median of the variation coefficients [41] calculated for each genome pair and its respective replicates. It was also evaluated to what extent the CI width was affected by the distribution of both numbers and sizes of HSPs derived from a respective pair of genomes. This was done for a selected, well performing BLAST+ method (see below). To assess the amount of uncertainty indicated by the bootstrap/jackknife CIs in DDH prediction, the bounds of each CI were transformed to DDH (in addition to the distance point estimate) using the investigated prediction model. Thus, the relationship between (i) the model’s CI estimating the uncertainty of DDH prediction, as detailed in the next section, and, (ii) the bootstrap/jackknife CI translated to DDH scale was assessed for each observation.
DDH prediction using sophisticated statistical models
The problems caused by linear models (see above) for predicting DDH via intergenomic distances can be solved by more sophisticated statistical models such as generalized linear models (GLMs) [42], generalized additive models (GAMs) [43] and Loess models [44] and identifying the one that provides the most robust predictions ([20], pp. 58-79). All models used DDH values as response variable and the corresponding intergenomic distance values as predictor variable. Among the novel models assessed in this study, GLMs came to our attention for several reasons: (i) they do not require the response variable to be normally distributed, (ii) the response predictions (here: DDH) would be bound within a fixed interval between a 0 and 100% and (iii) the predictor variable does not need to have a constant variance. The latter was especially important because, in our case, as usual for proportion data, the distance values are strictly defined in a range between 0 and 1, thus leading to a decrease in variance when approaching these boundaries, causing a dependency of the variance on the mean ([37], pp. 571).
GLMs belong to the parametric modeling techniques and make assumptions about the underlying distribution. For proportional response data as present here a binomial distribution is recommended ([37], pp. 515). One benefit of such a logistic regression is the variance-stabilizing effect on the response variable (removal of heteroscedasticity). DDH response data was appropriately converted to represent the number of failures and successes of an event ([37], pp. 574). Another special type of GLM was constructed by changing the response from DDH proportions to a binary response variable ([37], pp. 593-610). For any given intergenomic distance such a model yields the probability of whether or not it corresponds to a DDH value ≥ 70%. Finally, a non-parametric Loess smoother was also evaluated, as well as generalized additive models [45] but neither yielded better results than the GLMs (data not shown).
To assess whether the fit of the overall model (determined by the model’s residual deviance) could be further improved, a log transformation was applied to the explanatory variable [21, 22], and/or a variance-stabilizing arcsine transformation was applied to the response variable (see [37], pp. 570 and [35], pp. 386). In standard linear-regression models the coefficient of determination (R^{2}) provides a measure of how well future outcomes are likely to be predicted by a certain linear model. As GLMs do not provide R^{2} for model diagnosis, we checked for potential overdispersion (i.e., extra, unexplained variation) ([37], pp. 522) and where applicable used the Akaike information criterion [46] to measure the relative goodness of the GLM fits. Graphical evaluation of the model fits was done using the R package ggplot[47].
The performance of the model types and data transformations was also assessed by computing error ratios in DDH prediction. For each of the 4350 GBDP settings we calculated the models (under DS1) and compared the DDH predictions with the respective wet-lab DDH value. In a second pass, we calculated the model’s error ratio by assessing the number of false positives (i.e., a predicted DDH value equal or above 70% corresponding to a real DDH value below that threshold) and false negatives (i.e., a predicted DDH value below 70% corresponding to a real DDH value equal or above that threshold) relative to the total number of observations. To investigate the impact of the extension of the empirical data set, we chose the best-performing GBDP method from this study and fitted the aforementioned models to both the full data set (DS1) and the reduced one (DS2).
Results
Performance of methods and settings in mimicking wet-lab DDH
Preselection of well-performing GBDP methods from the correlation analysis
Correlations | Settings | |||||
---|---|---|---|---|---|---|
Dataset | Type | Estimate | Alignment tool or method | E-value filter | Algorithm | Formula |
DS1 | Kendall | -0.761 | BLAT | 10 | Coverage | d_{6}, d_{8} |
-0.752 | BLAST+ (WL46) | 10 | Coverage | d_{6}, d_{8} | ||
-0.677 | BLAST+ (WL46) | 10 | Coverage | d _{4} | ||
Pearson | -0.956 | BLAT | 10 | Greedy | d _{6} | |
-0.956 | BLAT | 10^{−2} | Trimming | d _{6} | ||
-0.946 | BLAST+ (WL38) | 10 | Coverage | d _{4} | ||
-0.935 | BLAST+ (WL46) | 10 | Coverage | d_{6}, d_{8} | ||
DS2 | Kendall | -0.763 | BLAT | 10 | Coverage | d_{6}, d_{8} |
Pearson | -0.954 | BLAT | 10 | Coverage | d_{6}, d_{8} | |
DS3 | Kendall | -0.783 | BLAST+ (WL38) | any | Coverage | d_{6}, d_{8} |
-0.717 | ANI | - | - | - | ||
Pearson | -0.980 | MUMmer (MR20) | - | Greedy | d_{0}, d_{6} | |
-0.973 | ANI | - | - | - | ||
DS4 | Kendall | -0.737 | BLAT | 10, 10^{−2} | Coverage | d_{6}, d_{8} |
-0.735 | BLAST+ (WL45) | any | Coverage | d_{6}, d_{8} | ||
-0.693 | Tetra | - | - | - | ||
-0.598 | ANIb | - | - | - | ||
-0.594 | ANIm | - | - | - | ||
Pearson | -0.957 | BLAT | 10^{−2} | Greedy | d _{6} | |
-0.904 | ANIm | - | - | - | ||
-0.703 | ANIb | - | - | - | ||
-0.693 | Tetra | - | - | - |
The most influential GBDP parameters were assessed via a multiple regression under two types of model selection (forward selection and backward variable elimination) with both resulting in the same full regression model. The latter contained the independent variables “alignment tool”, “distance algorithm” and “distance formula” and all possible interaction terms between them (Additional file 5). That is, only “e-value filter method” got eliminated as it could not explain a significant amount of variation in the data. This was confirmed by the relative importance index (see Additional file 6, pictures 1 and 2) and an ANOVA (here, p-values were 0.8136 for Pearson’s ρ as response variable and 0.6594 for Kendall’s τ). Additional file 6 (pictures 3-6) shows the interaction effects and the impact of all independent variables.
Confidence intervals via bootstrapping or jackknifing
The relationship between the intergenomic distance and the underlying set of HSPs obtained by comparing the respective pair of genomes is presented in Additional file 6 (picture 10). In brief, the lengths of the HSPs are more unevenly distributed for more similar genomes (smaller distance values).
Models for DDH prediction and species delineation
Error ratios of selected GBDP methods
Model types | ||||
---|---|---|---|---|
GBDP settings | GLM_{ log } | GLM | LM_{ AS } | LM |
BLAT (d_{6}) | 0.045 | 0.058 | 0.052 | 0.052 |
BLAT (d_{8}) | 0.045 | 0.090 | 0.097 | 0.084 |
BLAST+ (d_{6}) | 0.090 | 0.065 | 0.071 | 0.052 |
BLAST+ (d_{8}) | 0.090 | 0.213 | 0.187 | 0.316 |
BLAST+ (d_{4}) | 0.039 | 0.052 | 0.052 | 0.052 |
Discussion
Bootstrapping and jackknifing GBDP
With only minor differences between bootstrapping and jackknifing, the use of different algorithms had an obvious impact on the CIs of the resulting distances. The full implementation of the “coverage” algorithm allowed its application in connection with distance formulae d_{4} to d_{9} (so far it was restricted to d_{0} to d_{3}). Its CVs and, accordingly, its CIs and those of the resulting DDH estimates, were two orders of magnitude smaller than the ones resulting from the “greedy” and “greedy-with-trimming” algorithms (Figure 4 and Additional file 6, picture 9). The CIs of the DDH values estimated with the “coverage” algorithm showed a typical quality of proportion data ([37], p. 515), i.e., very high as well as very low distances between two genomes cause a reduced amount of variation between the resampled distances.
That the “greedy” and “greedy-with-trimming” algorithm yielded substantially higher CVs and CIs, as well as an increase of CVs and CIs with decreasing distance (and, thus, increasing DDH similarity) is most likely caused by the fact that here sets of HSPs, not genome positions are resampled. The observed Methanococcus genome pair confirms this, as its comparison yielded few very long HSPs with a high impact on the resulting distance values during resampling (Additional file 6, picture 10). Indeed, we observed the overall tendency that more similar genomes not only result in more HSPs but also in HSPs with much less equally distributed lengths. Hence, for “greedy” and “greedy-with-trimming”, the resampling of HSPs probably leads to an overestimation of the uncertainty in distance estimation and thus cannot be recommended over the resampling in conjunction with the “coverage” algorithm, which by construction is not depending on HSPs during the resampling phase. The relative sizes of the CIs also implies that those from “coverage” bootstrapping or jackknifing are by an order of magnitude smaller than the CIs from the DDH prediction via models, and that those from “greedy” and “greedy-with-trimming” resampling are by an order of magnitude higher than the model-based CIs.
For practical purposes this indicates that in conjunction with “coverage”, bootstrapping and jackknifing GBDP can safely be omitted because the uncertainty it indicates is always substantially lower than the one indicated by the CIs of the model. This result is as expected, because genome sequencing results in a large number of characters - actually, the largest possible number of characters that can be sampled from an organisms -, and, if appropriately calculated, the uncertainty of intergenomic distances should be much lower than the one of wet-lab DDH experiments. Experimental errors in determining DDH in the wet lab, however, are likely to be responsible for the uncertainties in fitting the models for predicting DDH in silico.
Models for DDH prediction
All previous studies [6-8] were based on simple linear regressions models (LMs) and did not consider data transformations. We observed that error ratios in DDH prediction, provided by LMs, are generally higher than those of the generalized linear models (GLMs). In addition to mere statistical parameters (such as overfitting, overdispersion, the AIC or R^{2}), the size of the error ratio in predicting whether DDH values are above or below 70% is of particular interest. As expected, it was revealed that the smaller the correlation between DDH and intergenomic distance, the higher the error ratio (see Additional file 6, picture 12).
Moreover, the GLMs combined with the log-trans- formed explanatory variable yielded a higher consis- tency between the correlation coefficients and the prediction success at the 70% boundary, and the better correlating GBDP methods had, on average, a lower error ratio if combined with these log-GLMs instead of any of the other models. Additionally, GLMs, if applied as shown, guarantee that even DDH predictions based on extreme distance (or similarity) values are between 0% and 100%. For obvious reasons, models for species delimitation should be as exact as possible and, thus, LMs here at least be considered as problematic. The overdispersion detected when diagnosing GLMs was presumably due to distinct pairs of strains sharing identical intergenomic distance values but at the same time showing distinct DDH values. This effect is called “unmodeled heterogeneity” and could also result from clustering of the DDH measurements ([48], p. 52-61), as observed in our data set. A switch to a “complementary log-log”-link function, as suggested in ([37], p. 594), didn’t further improve the model (data not shown).
The enlarged data set provided a globally increased significance of the inferred results. The comparison of selected GBDP methods applied to either the old [8] or the new data set, however, revealed only minor differences in the parameters estimated by the statistical models we tested.
Both theoretical and empirical results thus favor GLMs over standard linear-regression models for obtaining in-silico DDH replacement methods. Its improved DDH prediction capabilities offer GBDP as a quick and now even more reliable alternative to the DDH wet-lab technique, thus moving further forward within the transition process to a genome-based taxonomic gold standard.
The recommended GBDP method
In principle, multiple optimality criteria could be applied for selecting a GBDP variant that works best in DDH prediction, depending on the users’ priorities. The newly completed “coverage” algorithm, however, can unanimously be recommended, because it yielded the best correlations for both formula d_{6} (in general, and particularly combined with BLAST+ using a word length of 46 and no e-value filtering) and formula d_{4} (in general, and particularly combined with BLAST+ using a word length of 38 and no e-value filtering). When dealing with incomplete genomes it is highly recommended to use formula d_{4}, as it is independent of sequence length, and thus not directly affected by the removal of HSPs due to the removal of parts of the genome [8] (see also Additional file 3). Here, formula d_{4} resulted in worse Kendall correlations but better Pearson correlations and error ratios at the 70% boundary than d_{6}, as observed earlier [8].
Regarding local-alignment programs, only BLAT performed better than BLAST+ combined with optimized settings, and only slightly so. BLAST+ ’s optimal initial word length setting of 38 or 46 allows for comparatively quick genome-genome comparisons, because the intergenomic search space is significantly reduced compared to the default value of 11, resulting in a lower execution time. A higher initial word length results in a lowered sensitivity of the local-alignment program, which had a positive effect on the correlation outcome, as previously reported [8]. This is in agreement with the fact that BLAT, which is a considerably less sensitive alternative to BLAST, overall performed best [8]. All in all, we conclude that the default setting for the novel GBDP implementation should be BLAST+ combined with d_{4} and the accordingly optimized settings regarding word length and e-value filtering, and that the corresponding log-GLM model should be used for predicting DDH including CIs. Moreover, the situations in which a user might instead favor BLAT over BLAST+ and/or d_{6} over d_{4} are straightforward to identify. All these recommendations can now be directly utilized via our updated web service (GGDC 2.0) at http://ggdc.dsmz.de.
Beyond pairwise distances
Since the dawn of computer-based approaches to phylogenetics, researchers were trying to devise solutions for assessing statistical support of the inferred phylogenies [49-52]. Apparently, branches lacking sufficient support should not be overestimated regarding the explanation and visualization of evolutionary events. Particularly bootstrapping and jackknifing [53] are widely-used solutions for this kind of question and can be applied to both aligned molecular sequences (multiple sequence alignments) and matrices of phenotypic characters. Here, resampling is applied to the characters, usually present as columns of a matrix whose rows represent the organisms, phylogenetic inference is applied to the resampled matrix, and finally a majority-rule consensus tree is calculated from the trees from all replicates [51]. If distance methods for phylogenetic inference such as neighbour-joining [53] are used in such a scenario, within each replicate a distance matrix is computed from a character matrix that has been resampled at once.
In contrast, distance methods that avoid the construc- tion of a character matrix would need to apply boots- trapping or jackknifing to each pairwise comparison independently. For instance, [54] developed a method that relies on pairwise sequence alignment only; here, maximum-likelihood distances are inferred, and bootstrapped independently, from the alignments of all pairs of sequences. To highlight the conceptual difference, the procedure was called “pseudo-bootstrapping”, and it was demonstrated to be conservative compared to bootstrap analysis of multiple sequence alignments [54].
Apparently, GBDP’s new bootstrapping and jackknifing facilities would also yield such a pseudo-bootstrapping approach if it was applied to phylogenetic problems. In particular, sequence comparison is conducted independently for all genome pairs involved, and the sets of HSPs or the coverage vectors - on which each pairwise comparison is based - never form a common character matrix [12, 13]. Bootstrapping and/or jackknifing would just add the individual resampling of these independently constructed sets of HSPs or coverage vectors. An advantage to phylogenomics provided by GBDP-bootstrapping over supermatrix approaches (which concatenate alignments of individual orthologous genes; see [4] for an overview of phylogenomic methods in the context of microbial taxonomy) is that the calculation of bootstrapped or jackknifed distances could be done incrementally, and only the phylogenetic inference from all formed distance matrices would need to be done after each update of the set of organisms of interest.
For this reason, GBDP with resampling could be a faster and resource-saving alternative to more compute-intense phylogenomics methods, particularly because GBDP can as well be applied to sequences from proteomes [13]. Besides, it easily copes with various phylogenetic problems such as paralogous genes [12], low-complexity regions [13] and unbalanced genome/proteome sizes [8]. However, whereas this study already presented evidence that “coverage” should be preferred over “greedy” and “greedy-with-trimming” if coupled with bootstrapping or jackknifing, it is a partially open question whether, and under which conditions, resampling proteome-based GBDP[13] should be preferred over analyzing nucleotide sequences this way [12]. Even though it is likely that the deeper branches of the phylogeny can only be resolved based on amino acid sequences GBDP[13], in-depth comparisons of the performance of GBDP- bootstrapping/jackknifing with more common phylogenomics methods, as well as similar methods that are also based on resampling HSPs [55], are still needed.
Nevertheless, that a single method can be applied to both genome-based species delimitation and phylogenomic inferences at other taxonomic levels, and that it can be coupled with the assessment of statistical significance at one level, already strongly indicates that GBDP is an important tool in the transition process to genome-based gold standards at all taxonomic levels. A tighter coupling between phylogenetic inference and the assignment of taxonomic ranks might also help to overcome what we regard as the most severe theoretical limitation of the DDH 70% rule: that a taxon defined as all organisms whose similarity to a type organism is above a certain threshold is never guaranteed to form a monophyletic group [53]. The practical limitations of the wet lab-based DDH, however, already seem to have been overcome.
Conclusions
This update on the GBDP method is an important enhancement, not only because existing features of the software have been improved but particularly because novel features have been added. Whereas GBDP was already shown to yield better correlation results in DDH prediction than ANI [6] in an earlier correlation study [8], we can also confirm this with respect to the JSpecies[7] implementation. Since taxonomists generally consider these approaches as potential “next generation” replacements for the traditional and currently still dominating wet-lab method [3, 4], up to now these approaches could not be used to determine the CI of intergenomic distance measures, thus rendering the latest installment of the GBDP method to be the first one supporting that feature. This is crucial because numeric estimations from empirical data (such as wet-lab DDH values) always yield a certain degree of uncertainty, and it is thus commonplace in statistics to provide measures of variation and confidence.
By introducing (i) bootstrapping and jackknifing to the GBDP approach, (ii) better performing DDH prediction models and the CIs they provide, and, (iii) direct calculation of the probability that an intergenomic distance yielded a DDH larger than 70%, the here presented methods provide an attractive alternative to the wet-lab DDH for current taxonomic techniques. The addition of novel distance functions (by completing the implementation of the “coverage” distances) was also beneficial here, particularly in conjunction with the novel resampling techniques and with respect to the resulting correlations with DDH.
Declarations
Acknowledgements
Cordial thanks are addressed to Marek Dynowski and Werner Dilling, both Zentrum für Datenverarbeitung, University of Tübingen, for granting access and for their technical support related to the compute clusters of the bwGRiD [33]. The authors thank Lea A.I. Vaas, CBS, Utrecht, for helpful suggestions.
Authors’ Affiliations
References
- Wayne LG, Brenner DJ, Colwell RR, Grimont PaD, Kandler O, Krichevsky MI, Moore LH, Moore WEC, Murray RGE, Stackebrandt E, Starr MP, Truper HG: Report of the Ad Hoc committee on reconciliation of approaches to bacterial systematics. Int J Syst Bacteriol 1987,37(4):463-464. 10.1099/00207713-37-4-463View ArticleGoogle Scholar
- Stackebrandt E, Goebel BM: Taxonomic note: a place for DNA-DNA reassociation and 16S rRNA sequence analysis in the present species definition in bacteriology. Int J Syst Bacteriol 1994,44(4):846-849. 10.1099/00207713-44-4-846View ArticleGoogle Scholar
- Schleifer K: Classification of Bacteria and Archaea: past, present and future. Syst Appl Microbiol 2009,32(8):533-542. 10.1016/j.syapm.2009.09.002View ArticlePubMedGoogle Scholar
- Klenk HP, Göker M: En route to a genome-based classification of Archaea and Bacteria? Syst Appl Microbiol 2010,33(4):175-182. 10.1016/j.syapm.2010.03.003View ArticlePubMedGoogle Scholar
- Vandamme P, Pot B, Gillis M, de Vos P: Polyphasic taxonomy, a consensus approach to bacterial systematics. Microbiol Rev 1996,60(2):407-438.PubMed CentralPubMedGoogle Scholar
- Goris J, Konstantinidis K, Klappenbach J, Coenye T, Vandamme P, Tiedje J: DNA-DNA hybridization values and their relationship to whole-genome sequence similarities. Int J Syst Evol Microbiol 2007, 57: 81-91. 10.1099/ijs.0.64483-0View ArticlePubMedGoogle Scholar
- Richter M, Rossello R: Shifting the genomic gold standard for the prokaryotic species definition. Proc Nat Acad Sci 2009,106(45):19126-19131. 10.1073/pnas.0906412106PubMed CentralView ArticlePubMedGoogle Scholar
- Auch AF, von Jan M, Klenk HP, Göker M: Digital DNA-DNA hybridization for microbial species delineation by means of genome-to-genome sequence comparison. Stand Genomic Sci 2010, 2: 117-134. 10.4056/sigs.531120PubMed CentralView ArticlePubMedGoogle Scholar
- De Ley J, De Smedt J: Improvements of the membrane filter method for DNA:rRNA hybridization. Antonie van Leeuwenhoek 1975, 41: 287-307. 10.1007/BF02565064View ArticlePubMedGoogle Scholar
- Klenk HP, Haas B, Schwass V, Zillig W: Hybridization homology: a new parameter for the analysis of phylogenetic relations, demonstrated with the urkingdom of the archaebacteria. J Mol Evol 1986, 24: 167-173. 10.1007/BF02099964View ArticleGoogle Scholar
- Woese CR, Kandler O, Wheelis ML: Towards a natural system of organisms: proposal for the domains Archaea, Bacteria, and Eucarya. Proc Nat Acad Sci 1990,87(12):4576-4579. 10.1073/pnas.87.12.4576PubMed CentralView ArticlePubMedGoogle Scholar
- Henz S, Huson D, Auch AF, Nieselt-Struwe K, Schuster S: Whole-genome prokaryotic phylogeny. Bioinformatics 2005,21(10):2329-2335. 10.1093/bioinformatics/bth324View ArticlePubMedGoogle Scholar
- Auch AF, Henz S, Holland B, Göker M: Genome BLAST distance phylogenies inferred from whole plastid and whole mitochondrion genome sequences. BMC Bioinformatics 2006, 7: 350. 10.1186/1471-2105-7-350PubMed CentralView ArticlePubMedGoogle Scholar
- Auch AF, Henz SR, Göker M: Phylogenies from whole genomes: Methodological update within a distance-based framework. German conference on Bioinformatics, Tübingen 2006. Tübingen [http://nbn-resolving.de/urn:nbn:de:bsz:21-opus-34178] Tübingen []
- Auch AF: A phylogenetic potpourri - Computational methods for analysing genome-scale data. PhD thesis. Universität Tübingen, Wilhelmstr. 32, 72074 Tübingen 2009, [http://nbn-resolving.de/urn:nbn:de:bsz:21-opus-44779] Universität Tübingen, Wilhelmstr. 32, 72074 Tübingen 2009, []
- Auch AF, Klenk HP, Göker M: Standard operating procedure for calculating genome-to-genome distances based on high-scoring segment pairs. Stand Genomic Sci 2010, 2: 142-148. 10.4056/sigs.541628PubMed CentralView ArticlePubMedGoogle Scholar
- Altschul S, Gish W, Miller W, Myers E, Lipman D: Basic local alignment search tool. J Mol Biol 1990,215(3):403-410.View ArticlePubMedGoogle Scholar
- Saitou N, Nei M: The neighbor-joining method: a new method for reconstructing phylogenetic trees. Mol Biol Evol 1987,4(4):406-425.PubMedGoogle Scholar
- Göker M, Grimm GW, Auch AF, Aurahs R, Kučera M: A clustering optimization strategy for molecular taxonomy applied to planktonic foraminifera ssU rDnA. Evol Bioinf 2010, 6: 97-112.View ArticleGoogle Scholar
- Motulsky H, Christopoulos A: Fitting Models to Biological Data Using Linear and Nonlinear Regression: A Practical Guide to Curve Fitting. Oxford: Oxford University Press; 2004.Google Scholar
- Fletcher D, MacKenzie D, Villouta E: Modelling skewed data with many zeros: A simple approach combining ordinary and logistic regression. Environ Ecol Stat 2005, 12: 45-54. 10.1007/s10651-005-6817-1View ArticleGoogle Scholar
- Lin SM, Du P, Huber W, Kibbe Wa: Model-based variance-stabilizing transformation for Illumina microarray data. Nucleic acids Res 2008,36(2):e11.PubMed CentralView ArticlePubMedGoogle Scholar
- Efron B: Bootstrap methods: another look at the jackknife. Ann Stat 1979, 7: 1-26. 10.1214/aos/1176344552View ArticleGoogle Scholar
- Miller RG: The jackknife - a review. Biometrika 1974, 61: 1-15.Google Scholar
- Pagani I, Liolios K, Jansson J, Chen IMa, Kyrpides NC, Smirnova T: The Genomes OnLine Database (GOLD) v.4: status of genomic and metagenomic projects and their associated metadata. Nucleic acids Res 2012,40(Database issue):D571—D579.PubMed CentralPubMedGoogle Scholar
- Camacho C, Coulouris G, Avagyan V, Ma N, Papadopoulos J, Bealer K, Madden T: BLAST+: architecture and applications. BMC Bioinformatics 2009, 10: 421. 10.1186/1471-2105-10-421PubMed CentralView ArticlePubMedGoogle Scholar
- Korf I, Yandell M, Bedell J: BLAST. Sebastopol: O’Reilly Media; 2003.Google Scholar
- Legendre P, Legendre L: Numerical Ecology. Amsterdam: Elsevier; 1998.Google Scholar
- Kurtz S, Phillippy A, Delcher AL, Smoot M, Shumway M, Antonescu C, Salzberg SL: Versatile and open software for comparing large genomes. Genome Biol 2004,5(2):R12. 10.1186/gb-2004-5-2-r12PubMed CentralView ArticlePubMedGoogle Scholar
- Kent W: BLAT - the BLAST-like alignment tool. Genome Res 2002,12(4):656-664.PubMed CentralView ArticlePubMedGoogle Scholar
- Schwartz S, Kent WJ, Smit A, Zhang Z, Baertsch R, Hardison RC, Haussler D, Miller W: Human-mouse alignments with BLASTZ. Genome Res 2003, 13: 103-107. 10.1101/gr.809403PubMed CentralView ArticlePubMedGoogle Scholar
- Bader D, Pennington R: Cluster computing: applications. Int J High Perform Comput 2001,15(2):181-185. 10.1177/109434200101500211View ArticleGoogle Scholar
- BwGRiD: Member of the German D-Grid initiative, funded by the Ministry of Education and Research and the Ministry for Science, Research and Arts Baden-Wuerttemberg (2007-2012). Tech. rep. Universities of Baden-Württemberg 2012. [http://www.bw-grid.de/] Tech. rep. Universities of Baden-Württemberg 2012. []
- Meier-Kolthoff JP, Auch AF, Klenk HP, Göker M: GBDP on the grid: a genome-based approach for species delimitation adjusted for an automated and highly parallel processing of large data sets. In Hochleistungsrechnen in Baden-Württemberg - Ausgewählte Aktivitäten im bwGRiD 2012. Karlsruhe: KIT Scientific Publishing; Forthcoming 2013.Google Scholar
- Sokal R, Rohlf F: Biometry: The Principles and Practice of Statistics in Biological Research. San Francisco: W.H. Freeman and Company; 1969.Google Scholar
- R Development Core Team: R: a Language and Environment for Statistical computing. Vienna: R Foundation for Statistical Computing; 2011. [http://www.r-project.org] []Google Scholar
- Crawley MJ: The R book. Chichester: Wiley Publishing; 2007.View ArticleGoogle Scholar
- Venables WN, Ripley BD: Modern Applied Statistics with S. New York: Springer; 2002.View ArticleGoogle Scholar
- Fox J: Effect displays in {R} for generalised linear models. J Stat Software 2003,8(15):1-27.View ArticleGoogle Scholar
- Grömping U: Relative importance for linear regression in R: the package relaimpo. J Stat Software 2006, 17: 1-27. 10.1360/jos170001View ArticleGoogle Scholar
- Hendricks W, Robey K: The sampling distribution of the coefficient of variation. Ann Math Stat 1936,7(3):129-132. 10.1214/aoms/1177732503View ArticleGoogle Scholar
- Nelder JA, Wedderburn RWM: Generalized linear models. J R Stat Soc 1972,135(3):370-384.Google Scholar
- Hastie T, Tibshirani R: Generalized Additive Models. London: Chapman & Hall/CRC; 1990.Google Scholar
- Cleveland W: Robust locally weighted regression and smoothing scatterplots. J Am Stat Assoc 1979,74(368):829-836. 10.1080/01621459.1979.10481038View ArticleGoogle Scholar
- Wood SN: Fast stable restricted maximum likelihood and marginal likelihood estimation of semiparametric generalized linear models. J R Stat Soc (B) 2011, 73: 3-36. 10.1111/j.1467-9868.2010.00749.xView ArticleGoogle Scholar
- Akaike H: A new look at the statistical model identification. IEEE Trans Autom Control 1974,19(6):716-723. 10.1109/TAC.1974.1100705View ArticleGoogle Scholar
- Wickham H: Ggplot2: Elegant Graphics for Data Analysis. New York: Springer; 2009.View ArticleGoogle Scholar
- Hilbe J: Negative Binomial Regression. Cambridge: Cambridge Univ Pr; 2011.View ArticleGoogle Scholar
- Mueller LD, Ayala F J: Estimation and interpretation of genetic distance in empirical studies. Genetical Res 1982, 40: 127-137. 10.1017/S0016672300019005View ArticleGoogle Scholar
- Penny D, Hendy MD: Testing methods of evolutionary tree construction. Cladistics 1985,1(3):266-278. 10.1111/j.1096-0031.1985.tb00427.xView ArticleGoogle Scholar
- Felsenstein J: Confidence limits on phylogenies: an approach using the bootstrap. Evolution 1985,39(4):783-791. 10.2307/2408678View ArticleGoogle Scholar
- Penny D, Hendy M: Estimating the reliability of evolutionary trees. Mol Biol Evol 1986,3(5):403-417.PubMedGoogle Scholar
- Felsenstein J: Inferring Phylogenies. Sunderland: Sinauer Associates; 2004.Google Scholar
- Thorne J, Kishino H: Freeing phylogenies from artifacts of alignment. Mol Biol and Evol 1992,9(6):1148-1162.Google Scholar
- Clarke GDP, Beiko RG, Ragan MA, Charlebois RL: Inferring genome trees by using a filter to eliminate phylogenetically discordant sequences and a distance matrix based on mean normalized BLASTP scores. J Bacteriol 2002,184(8):2072-2080. 10.1128/JB.184.8.2072-2080.2002PubMed CentralView ArticlePubMedGoogle Scholar
Copyright
This article is published under license to BioMed Central Ltd. This is an Open Access article distributed under the terms of the Creative Commons Attribution License(http://creativecommons.org/licenses/by/2.0), which permits unrestricted use, distribution, and reproduction in any medium, provided the original work is properly cited.