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**|**Genome Biol**|**v.18; 2017**|**PMC5547491

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Genome Biol. 2017; 18: 150.

Published online 2017 August 7. doi: 10.1186/s13059-017-1272-5

PMCID: PMC5547491

Alan F. Rubin,^{1,}^{2,}^{3,}^{4} Hannah Gelman,^{4,}^{5} Nathan Lucas,^{6} Sandra M. Bajjalieh,^{6} Anthony T. Papenfuss,^{1,}^{2,}^{3,}^{7,}^{8} Terence P. Speed,^{1,}^{8} and Douglas M. Fowler^{}^{4,}^{9}

Douglas M. Fowler, Email: ude.wu@relwofd.

Received 2017 April 18; Accepted 2017 July 6.

Copyright © The Author(s). 2017

Deep mutational scanning is a widely used method for multiplex measurement of functional consequences of protein variants. We developed a new deep mutational scanning statistical model that generates error estimates for each measurement, capturing both sampling error and consistency between replicates. We apply our model to one novel and five published datasets comprising 243,732 variants and demonstrate its superiority in removing noisy variants and conducting hypothesis testing. Simulations show our model applies to scans based on cell growth or binding and handles common experimental errors. We implemented our model in Enrich2, software that can empower researchers analyzing deep mutational scanning data.

The online version of this article (doi:10.1186/s13059-017-1272-5) contains supplementary material, which is available to authorized users.

Exploring the relationship between sequence and function is fundamental to enhancing our understanding of biology, evolution, and genetically driven disease. Deep mutational scanning is a method that marries deep sequencing to selection among a large library of protein variants, measuring the functional consequences of hundreds of thousands of variants of a protein simultaneously. Deep mutational scanning has greatly enhanced our ability to probe the protein sequence-function relationship [1] and has become widely used [2]. For example, deep mutational scanning has been applied to comprehensive interpretation of variants found in disease-related human genes [3, 4], understanding protein evolution [5–9], and probing protein structure [10, 11] with many additional possibilities on the horizon [2].

In a deep mutational scan, a library of protein variants is first introduced into a model system [12]. Model systems that have been used in deep mutational scanning include phage, bacteria, yeast, and cultured mammalian cells. A selection is applied for protein function or another molecular property of interest, altering the frequency of each variant according to its functional capacity. Selections can be growth-based or implement physical separation of variants into bins, as in phage display or flow sorting of cells. Next, the frequency of each variant in each time point or bin is determined by using deep sequencing to count the number of times each variant appears. Here, the variable region is either directly sequenced using a single-end or paired-end strategy, or a short barcode that uniquely identifies each variant in the population is sequenced instead [12, 13]. Barcoding enables accurate assessment of variable regions longer than a single sequencing read [4, 13, 14]. Analysis of the change in each variant’s frequency throughout the selection yields a score that estimates the variant’s effect. Scoring the performance of individual variants is distinct from a related class of methods that quantify tolerance for change at each position in a target protein [15]. Those approaches enable a different set of biological inferences that we do not seek to address here. Guidelines for the design of deep mutational scanning experiments have been discussed elsewhere [12, 16–18].

Fundamental gaps remain in our ability to use deep mutational scanning data to accurately measure the effect of each variant because practitioners lack a unifying statistical framework within which to interpret their results. Existing methods are diverse in terms of their scoring function, statistical approach, and generalizability. Two established implementations of deep mutational scanning scoring methods, Enrich [19] and EMPIRIC [20], calculate variant scores based on the ratio of variant frequencies before and after selection. This type of ratio-based scoring has been used to quantify the effect of non-coding changes in promoters as well [21]. However, while intuitive and easy to calculate, ratio-based scores are highly sensitive to sampling error when frequencies are low. For experimental designs that sample from more than two time points to improve the resolution of changes in frequency, ratio-based scoring is insufficient so a regression-based approach has been used instead [4, 16, 22, 23]. Both ratio and regression analyses can incorporate corrections for wild-type performance [8, 16, 19, 20, 24] or nonsense variants [20, 22] at the expense of restricting the method to protein-coding targets only.

The lack of a common standard for calculating scores makes comparison between studies difficult and existing bespoke methods are not applicable to the diverse array of experimental designs currently being used. Furthermore, no existing method quantifies the uncertainty surrounding each score, which limits the utility of the data. For example, one of the most compelling applications of deep mutational scanning is to annotate variants found in human genomes with the goal of empowering variant interpretation [4], where estimation of the uncertainty associated with each measurement in a common framework is crucial. At best, current approaches employ ad hoc filtering of putative low-quality scores, often using manually determined read-depth cutoffs.

To address these limitations, we present Enrich2, an extensible and easy-to-use computational tool that implements a comprehensive statistical model for analyzing deep mutational scanning data. Enrich2 includes scoring methods applicable to deep mutational scans with any number of time points. Unlike existing methods, Enrich2 also estimates variant scores and standard errors that reflect both sampling error and consistency between replicates. We explore Enrich2 performance using novel and published deep mutational scanning datasets comprising 243,732 variants in five target proteins, as well as simulated data. We demonstrate that Enrich2’s scoring methods perform better than existing methods across multiple experimental designs. Enrich2 facilitates superior removal of noisy variants and improved detection of variants of small effect and enables statistically rigorous comparisons between variants. Enrich2 is platform-independent and includes a graphical interface designed to be accessible to experimental biologists with minimal bioinformatics experience.

We distilled the common features of a deep mutational scan into a generalized workflow (Fig. 1). After the experiment, each FASTQ file is quality filtered and variants are counted. For directly sequenced libraries, this involves calling the variant for each read (see “Methods”). For barcoded libraries, barcode counts are assigned to variants using an additional file that describes the many-to-one barcode-to-variant relationship. Next, the counts for each variant are normalized and a score is calculated that quantifies the change in frequency of each variant in each selection. Finally, each variant’s scores from replicate selections are combined into a single replicate score using a random-effects model. Variant standard errors are also calculated for each selection and replicate score, allowing the experimenter to remove noisy variants or perform hypothesis testing. Enrich2 is designed to enable users to implement other scoring functions, so long as they produce a score and a standard error. Thus, Enrich2 can serve as a framework for any counting-based enrichment/depletion experiment.

For experimental designs with three or more time points, Enrich2 calculates a score for each variant using weighted linear least squares regression. These time points can be variably spaced, as in samples from a yeast selection withdrawn at different times, or they can be uniformly spaced to represent rounds or bins, as in successive rounds of a phage selection. This method assumes the selection pressure is relatively constant during the course of the selection. Each variant’s score is defined as the slope of the regression line. For each time point in the selection, including the input time point, we calculate a log ratio of the variant’s frequency relative to the wild-type’s frequency in the same time point and regress these values on time. Regression weights are calculated for each variant in each time point based on the Poisson variance of the variant’s count (see “Methods”). We estimate a standard error for each score using the weighted mean square of the residuals about the fitted line. We calculate *p* values for each score using the *z*-distribution under the null hypothesis that the variant behaves like wild-type (i.e. has a slope of 0).

A problem with linear regression-based scoring is that the wild-type frequency often changes non-linearly over time in an experiment-specific and selection-specific manner (Fig. 2). Some linear model-based approaches subtract the wild-type score from each variant’s score [4, 22], ignoring this issue and potentially reducing score accuracy. A solution for this problem, which has been used extensively, is normalizing each variant’s score to wild-type at each time point [16, 20, 25–27]. We implemented per-time point normalization and compared variant standard errors calculated with and without wild-type normalization for a total of 14 replicates in three different experiments: a phage selection for BRCA1 E3 ubiquitin ligase activity; a yeast two-hybrid selection for BRCA1-BARD1 binding; and a phage selection for E4B E3 ubiquitin ligase activity (Table 1). In all cases, wild-type normalization resulted in significantly smaller variant standard errors (*p*≈0, binomial test, Additional file 1). Variants that remain non-linear after normalization are poorly fit by our regression model and have high standard errors. Thus, they can easily be identified for further examination or removal.

Wild-type frequency can change non-linearly. The change in frequency of the wild-type over the course of replicate selections is shown for (**a**) BRCA1 E3 ubiquitin ligase, (**b**) BRCA1-BARD1 binding, or (**c**) E4B E3 ubiquitin ligase. Each *colored line* represents **...**

Wild-type normalization is not always the best option. For example, some experimental designs do not have a wild-type sequence in the library, which precludes wild-type normalization. Furthermore, experiments subject to high levels of stochasticity arising from low read depth or limited sampling can benefit from normalization to the total number of reads rather than to wild-type [16]. Normalization to wild-type is also inappropriate in cases where the effect of the wild-type is incorrectly estimated or subject to high levels of error [16, 28]. To deal with these cases, Enrich2 also offers normalization using the number of reads instead of the wild-type count.

Wild-type non-linearity is not the only problem in scoring a typical selection. Each time point has a different number of reads per variant and time points with low coverage are more affected by sampling error. An example of this issue is found in one of the replicate selections for BRCA1 E3 ubiquitin ligase activity (Fig. 3a). To address this problem, Enrich2 downweights time points in the regression with low counts per variant. Without weighted regression, the experimenter is forced to choose between three undesirable options: using the low coverage time point and adding noise to the measurements; removing the time point and complicating efforts to compare replicates; or spending time and resources to re-sequence the time point. Weighting avoids these undesirable options, achieving lower variant standard errors as compared to ordinary regression (Fig. 3b). To show that this effect is general and not a feature of the specific BRCA1 replicate we analyzed, we downsampled reads from a single time point in the E4B E3 ubiquitin ligase dataset. We find that weighted regression reduces the mean standard error regardless of the fraction of reads removed (Fig. 3c, d). Finally, we show that weighted regression improves reproducibility between replicates in the BRCA1 E3 ubiquitin ligase dataset, even in the absence of any filtering (Fig. 3e, f). A previously developed Bayesian MCMC approach could be used to generate a posterior variance, which would be of similar value to our standard errors [28]. However, this approach would be impracticably slow for tens of thousands of variants.

Weighted least squares regression reduces standard error and improves replicate correlation. **a** The number of reads (*shaded blue bars*) and the distribution of variant regression weights (*boxplots*, *solid green line* is the median, *dotted green line* is the **...**

For experiments with only two sequenced populations or time points (e.g. “input” and “selected”), Enrich2 calculates the slope between the two time point log ratios, which is equivalent to frequently used ratio-based scoring methods [1, 19, 20, 24]. Unlike previous implementations of ratio-based scoring, we provide standard error estimates for each score using Poisson assumptions (see “Methods”).

Deep mutational scans are affected by various sources of error in addition to sampling error. One way to deal with this problem is to perform replicates. Usually, each variant’s score is calculated by taking the mean across replicates, which ignores the distribution of replicate scores. Furthermore, if an error is calculated, it is derived only from the replicate scores’ distribution and ignores any error associated with each replicate score. One alternative is to combine replicate scores using a fixed-effect model [29]. We examined this approach for the BRCA1 E3 ubiquitin ligase dataset (Fig. 4) and found that because variant scores can vary widely between replicates, this method dramatically underestimates the standard error of the combined variant score. We therefore implemented a random-effects model that estimates each variant’s score based on the distribution of that variant’s scores across all replicates. This random-effects model also produces a standard error estimate for each variant that captures selection-specific error as well as error arising from the distribution of replicate scores (see “Methods”).

A random-effects model for scoring replicate selections. Variant scores for 20 randomly selected variants from the BRCA1 E3 ubiqutin ligase dataset are shown. The replicate scores (*green*) were determined for each variant using Enrich2 weighted regression. **...**

The random-effects model furnishes variant scores that are less sensitive to outlier replicates than a fixed-effect model (Fig. 4). Additionally, standard errors estimated by the random-effects model better reflect the distribution of replicate scores, providing a better basis for subsequent hypothesis testing. The same random-effects model can be used for experiments with any number of time points or replicates or with any Enrich2 scoring function (Additional file 2: Figure S1). A key advantage of this approach is that error is quantified on a per-variant basis, unlike the usual approach of comparing replicate selections using pairwise correlation [4, 15, 22]. This allows experimenters to use replicate data to make inferences about individual variants, rather than simply as a quality control check for whole experiments.

Per-variant standard error estimates enable the removal of variants with unreliable scores. This contrasts with previous filtering schemes, which employed an empirical cutoff for the minimum number of read counts for each variant in the input library or throughout the selection [1, 4, 12, 14, 30–36]. Read count cutoffs eliminate low-count variants that may be unreliably scored due to sampling error, but ignore other sources of noise and may introduce a bias against variants that become depleted after selection. Enrich2 retains low-count variants and enables the experimenter to determine which scores are reliable directly from the associated standard error.

To assess whether standard error-based filtering performs better than read count-based filtering, we analyzed data from a deep mutational scan of the C2 domain of Phospholipase A2 (Table 1). Here, a library of 84,252 phage-displayed C2 domain variants was selected for lipid binding over several rounds. This dataset was un-analyzable using previous methods due to the apparent extreme variability between replicate selections. We compared filtering based on four different parameters: variant standard error calculated using the random-effects model or the fixed-effect model, read count in the input round, and total read count in all rounds of selection. To quantify filtering method performance, we took the top quartile of variants selected by each filtering method. Then, we calculated the pairwise Pearson correlation coefficient between variant scores for each possible pair of the three replicates in the C2 domain dataset (Fig. 5, Additional file 3). We found that filtering based on standard errors from the random-effects model was the only method that recovered a replicable subset of variants from this dataset. In fact, input count filtering selected a subset of variants whose scores were more poorly correlated than the unfiltered set. We performed a similar analysis on the higher-quality E4B, neuraminidase, and BRCA1 replicate datasets using the top three quartiles of variants. As for the C2 domain, we found that filtering based on random-effects standard error outperforms the other filtering methods (Additional file 3). For example, in the E4B dataset random-effects standard error filtering performed better (pairwise Pearson r^{2}=0.80) than fixed-effect standard error (r^{2}=0.59), input library count (r^{2}=0.58), or total count filtering (r^{2}=0.59). We note that any filtering strategy removes variants and reduces coverage. To explore how the stringency of variant filtering affects replicate correlation, we calculated replicate correlations after removing increasing numbers of variants according to each of the four filtering methods (Additional file 2: Figure S2). We found that filtering by standard errors from the random-effects model was the only approach that yielded high correlations between replicates for the C2 domain data. Furthermore, random-effects standard error filtering performed better at nearly all filtering stringencies in both the C2 domain and BRCA1 E3 datasets.

Standard error-based filtering improves replicate correlation. Variant scores from two replicates of the C2 domain dataset are shown. Each *panel* plots the top quartile of variants selected by standard error from the random-effects model (*leftmost column* **...**

To further demonstrate the utility of Enrich2 standard error-based filtering, we re-analyzed a deep mutational scan of the influenza virus neuraminidase gene (Table 1). In this experiment, 22 neuraminidase variants were individually validated and used to assess the quality of the deep mutational scanning data. Of these individually validated variants, four had large variant score standard errors as determined by Enrich2’s random-effects model (Fig. 6a, Additional file 2: Figure S3, Additional file 4). Removing these high-standard error variants improved the correlation between the deep mutational scanning scores and individual validation scores from Pearson r^{2}=0.81 to r^{2}=0.87. Removal of these variants also improved the correlation when scores were calculated as originally described in the study (Pearson r^{2}=0.80 versus r^{2}=0.84) (Additional file 2: Figure S3) [30]. This suggests that scores of variants with low Enrich2 standard errors are more likely to reflect the results of gold standard validation experiments and supports the use of standard error-based filtering for selecting candidate variants for follow-up studies. We note that in the neuraminidase experiment, the three replicates used a common starting library. This design fails to capture some artifacts, especially those introduced during cloning. Ideally, full biological replicates should be collected.

An important challenge in analyzing deep mutational scanning data is determining whether a variant behaves differently from wild-type or differently under altered conditions. Enrich2 standard errors empower experimenters to perform statistical tests for such differences. By default, Enrich2 calculates raw *p* values for each score under the null hypothesis that the variant’s score is indistinguishable from wild-type using a *z*-test. This allows the user to discriminate between variants with extreme scores due to sampling error or other noise from those that are confidently estimated to be different from wild-type. We note that Enrich2 provides raw *p* values and users should correct for multiple testing using their preferred method.

We can also use a *z*-test to determine whether variants have different functional consequences under altered experimental conditions. For example, deep mutational scans of the neuraminidase gene were conducted in the presence and absence of the small molecule neuraminidase inhibitor oseltamivir (Table 1). The original study identified five “drug-adaptive” variants, defined as those that outperformed wild-type in the presence of oseltamivir [30]. These five drug-adaptive variants included three known oseltamivir-resistant variants. In our reanalysis, we identified 22 drug-adaptive variants including all five variants found in the original study (Fig. 6b, Additional file 5). Fifteen of these 22 drug-adaptive variants also had a significantly higher score than wild-type in the absence of the inhibitor and therefore might be more likely to occur in natural virus populations. Our results agree broadly with the original analysis. By using Enrich2 to calculate scores and standard errors for variants across replicates, we were able to identify additional candidate variants with small but statistically significant effects, some of which could be of biological interest. Of course, any new candidate variants could be false positives and they would need to be individually validated, as was done in the original study.

Our analyses of experimental data suggest that Enrich2 is a useful tool for exploring and understanding deep mutational scanning datasets. In support of this, we generated simulated datasets with predetermined variant effects and compared mathematically predicted Enrich2 variant effect scores to scores calculated from simulated data. Using this approach, we demonstrate that the Enrich2 method can be applied to data from either cell growth or binding assays and can handle different types of noise.

Deep mutational scanning datasets can be generated using different selection assays. Nearly all scans employ either cell growth assays or binding assays, which are typically conducted using phage or yeast display [12]. To demonstrate that the Enrich2 method can meaningfully assign variant scores for both assay types, we simulated data where each variant’s true effect was predetermined (see “Methods”). In growth simulations, a variant’s true effect was the growth rate of a cell carrying that variant; in binding simulations, a variant’s true effect was the probability of a cell or phage carrying that variant progressing to the subsequent round of selection. In our simulations, each variant’s true effect was drawn from a normal distribution with the wild-type true effect in the 75th percentile of the distribution, which is consistent with empirical datasets (Additional file 6).

Each simulated dataset contained 10,000 unique variants including wild-type. For each selection, a starting variant population was independently generated and then five rounds of growth or binding selection were performed (see “Methods”). Five replicate selections were simulated for each dataset. Sequencing was simulated such that each variant had, on average, 200 reads. The resulting datasets were scored by Enrich2 using the weighted least squares regression method and replicates were combined using the random-effects model. We found that the Enrich2 scores are strongly correlated with predicted scores based on the true variant effects (r^{2}=0.995 for binding and r^{2}=0.992 for growth) (Fig. 7a). Thus, the Enrich2 method captures true variant effects for both growth-based and binding-based assays. We note that the relationship of these variant effects to a physical parameter of interest (e.g. *K*
_{d} for binding) depends on the specific conditions of the experiment [37–39].

Variant scoring for growth and binding experiments using simulated data. **a** Enrich2 variant effect scores derived from simulated data are plotted against expected Enrich2 scores based on true variant effects in the simulation. Enrich2 accurately scores **...**

We also simulated noisy data and evaluated Enrich2’s ability to identify affected variants. One type of noise is inconsistent variant effects between replicates, which can arise from cloning errors or experimental variation. We simulated datasets in which 2% of variants in each of the five biological replicates were randomly assigned a new true effect. As expected, noisy variants have higher standard errors than other variants (Fig. 7b) and standard error-based filtering is an effective tool for removing them (Fig. 7c). The magnitude of a noisy variant’s standard error is proportional to the magnitude of the difference between the variant’s original true effect and the resampled true effect (r^{2}=0.85 for binding and r^{2}=0.93 for growth; Additional file 2: Figure S4). Another type of noise arises from unexpected amplification or depletion of variant counts in a single time point, which can be due to polymerase chain reaction (PCR) jackpotting or other artifacts during the DNA isolation, amplification, and sequencing steps. We simulated datasets in which 10% of variants are over-represented or under-represented in a single time point. We found that the random-effects model accurately assigns scores to these amplified or depleted variants (Additional file 2: Figure S5A) and the affected variants are easily identified by their replicate standard errors (Additional file 2: Figure S5B). These results illustrate that the Enrich2 method is robust to common types of noise present in deep mutational scanning data.

We developed a statistical framework for analyzing deep mutational scanning data that is applicable to many common experimental designs. We showed that our statistical method is superior to existing methods for removing noisy variants and detecting variants of small effect, enabling researchers to extract more from their datasets. We implemented our method in Enrich2, a computationally efficient graphical software package intended to improve access to deep mutational scanning for labs without data analysis experience. Enrich2 is extensible, so users can implement and easily share new scoring functions as new deep mutational scanning experimental designs are developed.

Enrich2 builds upon previous approaches to regression-based scoring, which we improved in two ways. First, per-time-point wild-type normalization helps reduce the effects of non-linear behavior under the assumption that many sources of non-linearity affect most variants similarly. Second, weighting each regression time point based on variant counts helps alleviate sampling error. In addition to these improvements, Enrich2 combines replicate selections into a single set of variant scores with standard errors to help identify variants that behave consistently in a given assay. Though variant score precision does not guarantee accuracy, we showed that removing variants with high standard errors from the neuraminidase dataset did improve the correlation between deep mutational scanning results and gold-standard measurements.

Enrich2 furnishes generalized variant effect scores, which we showed are applicable to both growth-based and binding-based deep mutational scans. In the case of growth-based deep mutational scans, variant scores are linearly related to growth rate. In the case of binding-based deep mutational scans, variant scores are linearly related to the log of the likelihood of selection in each round. We note that the relationship between the likelihood of selection and variant binding affinity depends on experimental specifics including the number of molecules displayed per cell or phage, ligand concentration, and degree of non-specific binding [39]. Furthermore, the regression-based approach described here is designed for deep mutational scans with constant selection pressure. Selections conducted over longer timescales or selections in which the selection pressure is modulated by the experimenter may not be modeled accurately by our approach [8, 40, 41]. Specific scoring methods that take into account experimental details such as ligand concentration or variable selection pressure could easily be added to Enrich2, taking advantage of the program’s existing read counting, variant calling, replicate combining, and visualization machinery.

Enrich2 standard errors can also be used to conduct hypothesis tests comparing variants within a single experimental condition or between multiple conditions. When comparing variants between conditions, we assume that the distribution of scores between conditions is roughly similar, but this assumption does not hold in all cases. For example, the shape of the score distribution is a function of the strength of the selective pressure applied [8] and, more generally, the experimental conditions employed. Thus, Enrich2 standard errors should be used with caution when comparing variants between differing selections unless the variant scores are similarly distributed and the selection conditions are comparable. A general method for normalizing scores to facilitate comparisons across different conditions or selection pressures remains an important open question, as existing approaches are computationally intensive [28].

The use of deep mutational scanning is expanding rapidly and better tools for analysis will help it flourish. As with other widely used high-throughput experimental methods, a robustly implemented common statistical framework reduces barriers to entry, ensures data quality, and enables comparative analyses. We suggest that Enrich2 can help deep mutational scanning continue to grow by providing a foundation for meeting these challenges and facilitating further exploration and collaboration.

Enrich2 implements alignment-free variant calling. Variant sequences are expected to have the same length and start point as the user-supplied wild-type sequence, which allows Enrich2 to compare each variant to the wild-type sequence in a computationally efficient manner. In addition to this alignment-free mode, an implementation of the Needleman-Wunsch global alignment algorithm [42] is included that will call insertion and deletion events. Enrich2 supports overlapping paired-end reads and single-end reads for direct variant sequencing, as well as barcode sequencing for barcode-tagged variants.

For selections with at least three time points, we define *T*, which includes all time points, and *T*′, which includes all time points except the input (*t*
_{0}). The frequency of a variant (or barcode) *v* in time point *t* is the count of the variant in the time point (*c*
_{v,t}) divided by the number of reads sequenced in the time point (*N*
_{t}).

$${f}_{v,t}=\frac{{c}_{v,t}}{{N}_{t}}$$

The change in frequency for a variant *v* in a non-input time point *t* *T*′ is the ratio of frequencies for *t* and the input.

$${r}_{v,t}=\frac{{f}_{v,t}}{{f}_{v,0}}$$

Instead of using this raw change in variant frequency, we divide each variant’s ratio by the wild-type (*wt*) variant’s ratio.

$$\frac{{r}_{v,t}}{{r}_{\mathit{wt},t}}=\frac{{c}_{v,t}{c}_{\mathit{wt},0}}{{c}_{v,0}{c}_{\mathit{wt},t}}$$

Because the library size terms (*N*
_{t} and *N*
_{0}) in the frequencies cancel out, the ratio of ratios is not dependent on other non-wild-type variants in the selection. In practice, we add ½ to each count to assist with very small counts [43] and take the natural log of this ratio of ratios.

$${L}_{v,t}=log\left(\frac{\left({c}_{v,t}+\frac{1}{2}\right)\left({c}_{\mathit{wt},0}+\frac{1}{2}\right)}{\left({c}_{v,0}+\frac{1}{2}\right)\left({c}_{\mathit{wt},t}+\frac{1}{2}\right)}\right)$$

This equation can be rewritten as

$${L}_{v,t}=log\left(\frac{{c}_{v,t}+\frac{1}{2}}{{c}_{\mathit{wt},t}+\frac{1}{2}}\right)-log\left(\frac{{c}_{v,0}+\frac{1}{2}}{{c}_{\mathit{wt},0}+\frac{1}{2}}\right)$$

If we were to regress *L*
_{v,t} on *t* *T*
^{′}, we note that the second term is shared between all the time points and therefore only affects the intercept of the regression line. We do not use the intercept in the score, so instead we regress on *M*
_{v,t} and use all values of *t* *T*.

$${M}_{v,t}=log\left(\frac{{c}_{v,t}+\frac{1}{2}}{{c}_{\mathit{wt},t}+\frac{1}{2}}\right)$$

The score is defined as the slope of the regression line, ${\hat{\beta}}_{v}$. In practice, we regress on $\frac{t}{maxT}$ to facilitate comparisons between selections with different magnitudes of time points (e.g. 0/1/2/3 rounds versus 0/24/48/72 hours).

To account for unequal information content across time points with variable sequencing coverage, we perform weighted linear least squares regression [44]. The regression weight for *M*
_{v,t} is *V*
_{v,t}
^{−1}, where *V*
_{v,t} is the variance of *M*
_{v,t} based on Poisson assumptions [43] and is approximately

$${V}_{v,t}=\frac{1}{{c}_{v,t}+\frac{1}{2}}+\frac{1}{{c}_{\mathit{wt},t}+\frac{1}{2}}$$

For selections with only two time points (e.g. input and selected), we use the slope of the line connecting the two points as the score. This is equivalent to the wild-type adjusted log ratio (*L*
_{v}) derived similarly to *L*
_{v,t} above.

$${L}_{v}=log\left(\frac{{c}_{v,sel}+\frac{1}{2}}{{c}_{\mathit{wt},sel}+\frac{1}{2}}\right)-log\left(\frac{{c}_{v,inp}+\frac{1}{2}}{{c}_{\mathit{wt},inp}+\frac{1}{2}}\right)$$

As there is no residual error about the fitted line, we must use a different method to estimate the standard error. We calculate a standard error (*SE*
_{v}) for the enrichment score *L*
_{v} under Poisson assumptions [24, 43].

$$S{E}_{v}=\sqrt{\frac{1}{{c}_{\mathit{v,inp}}+\frac{1}{2}}+\frac{1}{{c}_{\mathit{wt,inp}}+\frac{1}{2}}+\frac{1}{{c}_{\mathit{v,sel}}+\frac{1}{2}}+\frac{1}{{c}_{\mathit{wt,sel}}+\frac{1}{2}}}$$

For experiments with no wild-type sequence, scores can be calculated using the filtered library size for each time point *t*, which is defined as the sum of counts at time *t* for variants that are present in all time points.

To account for replicate heterogeneity, we use a simple meta-analysis model with a single random effect to combine scores from each of the *n* replicate selections into a single score for each variant. Each variant’s score is calculated independently. Enrich2 computes the restricted maximum likelihood estimates for the variant score ($\hat{\beta}$) and standard error (${\hat{\sigma}}_{s}$) using Fisher scoring iterations [45]. Given the replicate scores (${\hat{\beta}}_{i}$) and estimated standard errors (${\hat{\sigma}}_{i}$) where *i*=1, 2, …, *n*, the estimate for $\hat{\beta}$ at each iteration is the weighted average:

$$\widehat{\beta}=\frac{{\displaystyle {\sum}_{i=1}^{n}}{\widehat{\beta}}_{i}{\left({\widehat{\sigma}}_{s}^{2}+{\widehat{\sigma}}_{i}^{2}\right)}^{-1}}{{\displaystyle {\sum}_{i=1}^{n}}{\left({\widehat{\sigma}}_{s}^{2}+{\widehat{\sigma}}_{i}^{2}\right)}^{-1}}$$

The starting value for ${\widehat{\sigma}}_{s}^{2}$ at the first iteration is:

$${\widehat{\sigma}}_{s}^{2}=\frac{1}{n-1}{\displaystyle \sum _{i=1}^{n}}{\left({\widehat{\beta}}_{i}-\overline{\widehat{\beta}}\right)}^{2}$$

Enrich2 calculates the following fixed-point solution for ${\widehat{\sigma}}_{s+1}^{2}$:

$${\widehat{\sigma}}_{s+1}^{2}={\widehat{\sigma}}_{s}^{2}\frac{{\displaystyle {\sum}_{i=1}^{n}}{\left({\widehat{\sigma}}_{s}^{2}+{\widehat{\sigma}}_{i}^{2}\right)}^{-2}{\left({\widehat{\beta}}_{i}-\widehat{\beta}\right)}^{2}}{{\displaystyle {\sum}_{i=1}^{n}}{\left({\widehat{\sigma}}_{s}^{2}+{\widehat{\sigma}}_{i}^{2}\right)}^{-1}-\frac{{\displaystyle {\sum}_{i=1}^{n}}{\left({\widehat{\sigma}}_{s}^{2}+{\widehat{\sigma}}_{i}^{2}\right)}^{-2}}{{\displaystyle {\sum}_{i=1}^{n}}{\left({\widehat{\sigma}}_{s}^{2}+{\widehat{\sigma}}_{i}^{2}\right)}^{-1}}}$$

Because it is more computationally efficient to perform a fixed number of iterations for all variant scores in parallel than to test for convergence of each variant, Enrich2 performs 50 Fisher scoring iterations. In practice, this is more than sufficient for ${\widehat{\sigma}}_{s}^{2}$ to converge. We record the difference ${\epsilon}_{s}={\widehat{\sigma}}_{s}^{2}-{\widehat{\sigma}}_{s-1}^{2}$ for the final iteration and identify any variants with high values for
_{s} as variants that failed to converge. No such variants were encountered in the analyses detailed here.

For the fixed-effect model [29], we calculate the variant score (${\hat{\beta}}^{\mathit{\text{'}}}$) and standard error (${\widehat{\sigma}}_{s}^{\mathit{\text{'}}}$) using a weighted average of the replicate scores (${\hat{\beta}}_{i}$) where the weight for each score is the inverse of that variant’s variance (${\widehat{\sigma}}_{s-1}^{2}$). The standard error of the variant ${\widehat{\sigma}}_{s}^{\mathit{\text{'}}}$ is:

$${\widehat{\sigma}}_{s}^{\text{'}}=\sqrt{\frac{1}{{\displaystyle {\sum}_{i=1}^{n}}{\widehat{\sigma}}_{i}^{-2}}}$$

The fixed-effect model was used for comparison purposes only and is not implemented in the Enrich2 software.

The behavior of a variant *v* in a simulated binding experiment (e.g. phage display, yeast display) can be described in terms of the displaying entity’s likelihood of being selected in a given round [39, 46]. This likelihood is related to the binding affinity of each variant, and, by extension, the binding probability of an individual protein molecule under the experimental conditions. The relationship between variant binding affinity, monomer binding probability, and likelihood of selection will depend on the specifics of the experiment such as the number of molecules displayed per cell or phage, ligand concentration, and non-specific binding [39]. Each round of selection is a time point *t* in the analysis, so we can assign each variant a probability of being selected in a given time point (*p*
_{v,t}). We assume that *p*
_{v,t}=*p*
_{v,0}=*p*
_{v} (i.e. that the probability is constant throughout the selection) and that any grow out or amplification steps are uniform across all variants.

The initial variant population is determined by the variant population frequencies (${f}_{v,0}^{\mathit{\prime}}$) and the size of the starting population (${N}_{0}^{\prime}$).

$${c}_{v,\phantom{\rule{0.25em}{0ex}}0}^{\prime}={f}_{v,\phantom{\rule{0.25em}{0ex}}0}^{\prime}{N}_{0}^{\prime}$$

We note that ${c}_{v,\mathrm{t}}^{\mathit{\prime}}$, ${f}_{v,\mathrm{t}}^{\mathit{\prime}}$, and ${N}_{\mathrm{t}}^{\prime}$ refer to the variant population itself, in contrast to the previously defined *c*
_{v,t}, *f*
_{v,t}, and *N*
_{t}, which refer to sequence reads derived from the variant population.

We define *a*
_{t} as a factor describing growth between round *t* and the previous round (*a*
_{0}=1). We assume that *a*
_{t} is the same for all variants. The count for a variant in time point *t*+1 in terms the count in time point *t* is:

$${c}_{v,\phantom{\rule{0.25em}{0ex}}t+1}^{\prime}={a}_{t+1}{p}_{v}{c}_{v,\phantom{\rule{0.25em}{0ex}}t}^{\prime}$$

Therefore, the count for a variant in time point *t* given the starting count is:

$${c}_{v,t}^{\text{'}}={c}_{v,0}^{\text{'}}{\displaystyle \prod _{j=1}^{t}}{a}_{j}{p}_{v}={f}_{v,0}^{\text{'}}{N}_{0}^{\text{'}}{p}_{v}^{t}{\displaystyle \prod _{j=1}^{t}}{a}_{j}$$

We can write the ratio of variant counts in these terms and define the log ratio for binding experiments (*M*
_{v,t}
^{'}).

$$\frac{{c}_{v,t}^{\text{'}}}{{c}_{\mathit{wt},t}^{\text{'}}}=\frac{{f}_{v,0}^{\text{'}}{N}_{0}^{\text{'}}\phantom{\rule{0.1em}{0ex}}{p}_{v}^{t}{\displaystyle {\prod}_{j=1}^{t}}{a}_{j}}{{f}_{\mathit{wt},0}^{\text{'}}{N}_{0}^{\text{'}}\phantom{\rule{0.5em}{0ex}}{p}_{\mathit{wt}}^{t}{\displaystyle {\prod}_{j=1}^{t}}{a}_{j}}=\frac{{f}_{v,0\phantom{\rule{0.1em}{0ex}}}^{\text{'}}{p}_{v}^{t}}{{f}_{\mathit{wt},0\phantom{\rule{0.1em}{0ex}}}^{\text{'}}{p}_{\mathit{wt}}^{t}}$$

$${M}_{v,t}^{\text{'}}=log\left(\frac{{c}_{v,t}^{\text{'}}}{{c}_{\mathit{wt},t}^{\text{'}}}\right)=t\cdot log\left(\frac{{p}_{v}}{{p}_{\mathit{wt}}}\right)+log\left(\frac{{f}_{v,0}^{\text{'}}}{{f}_{\mathit{wt},0}^{\text{'}}}\right)$$

If we substitute *t* for ${t}^{\prime}=\frac{t}{max\phantom{\rule{0.3em}{0ex}}T}$, we find that the expected score for binding experiments under the regression scoring model (${\beta}_{v}^{\prime}$) should be related to the variant selection probability (*p*
_{v}) by:

$${\beta}_{v}^{\prime}=(max\phantom{\rule{0.3em}{0ex}}T)\mathrm{log}\left(\frac{{p}_{v}}{{p}_{wt}}\right)$$

The behavior of a variant *v* in a simulated growth experiment can be described by the growth rate at time *t* (*μ*
_{v}(*t*)). Unlike in the round-based binding experiment case, time in growth experiments is modeled as continuous. We assume that *μ*
_{v}(*t*)=*μ*
_{v}(0)=*μ*
_{v} (i.e. that the growth rate is constant throughout the selection) and that any amplification steps are uniform across all variants. This derivation is based on [16, 18]. In interference-free growth, the growth of individual variants can be described by the first order equation:

$$\frac{d{c}_{\mathit{v}}^{\text{'}}}{dt}={\mu}_{v}{c}_{v}^{\text{'}}(t)$$

Therefore, the count for a variant at time *t* given the starting count is:

$${c}_{v}^{\text{'}}(t)={c}_{v,0}^{\text{'}}{e}^{{\mu}_{v}t}={f}_{v,0}^{\text{'}}{N}_{0}^{\text{'}}{e}^{{\mu}_{v}t}$$

We can write the ratio of variant counts in these terms and construct the continuous function *M*
_{v}
^{″}(*t*).

$$\frac{{c}_{v}^{\text{'}}\left(t\right)}{{c}_{\mathit{wt}}^{\text{'}}\left(t\right)}=\phantom{\rule{0.25em}{0ex}}\frac{{N}_{0}^{\text{'}}\phantom{\rule{0.2em}{0ex}}{f}_{v,0}^{\text{'}}{e}^{{\mu}_{v}t}}{{N}_{0}^{\text{'}}\phantom{\rule{0.2em}{0ex}}{f}_{\mathit{wt},0}^{\text{'}}{e}^{{\mu}_{\mathit{wt}}t}}\phantom{\rule{0.2em}{0ex}}=\phantom{\rule{0.25em}{0ex}}\frac{{f}_{v,0}^{\text{'}}}{{f}_{\mathit{wt},0}^{\text{'}}}{e}^{\left({\mu}_{v}-{\mu}_{\mathit{wt}}\right)t}$$

$${M}_{v}^{\u2033}\left(t\right)=log\left(\frac{{c}_{v}^{\text{'}}\left(t\right)}{{c}_{\mathit{wt}}^{\text{'}}\left(t\right)}\right)=\left({\mu}_{v}-{\mu}_{\mathit{wt}}\right)t+log\left(\frac{{f}_{v,0}^{\text{'}}}{{f}_{\mathit{wt},0}^{\text{'}}}\right)$$

We convert to the discrete function *M*
_{v,t}
^{″} for convenience by assuming that *m* timepoints are sampled at constant intervals, determined by the number of wild-type doublings (*δ*) per time point, such that max *T*=*mδ*. We then find that the expected score for growth experiments under the regression scoring model (*β*
_{v}
^{″}) should be related to the growth rate (*μ*
_{v}) by:

$${\beta}_{v}^{\u2033}=m\delta \left({\mu}_{v}-{\mu}_{\mathit{wt}}\right)$$

Simulated datasets contain 10,000 unique variants (including wild-type), each characterized by a true variant effect: the probability of selection in each round (*p*
_{v}) for binding simulations or the growth rate (*μ*
_{v}) for growth simulations. We assume that the variant effect distribution is normal and set the wild-type effect to *p*
_{wt}=0.5 and *μ*
_{wt}=1. We set the wild-type effect at the 75th percentile of the distribution and set the standard deviation to 0.1. We draw 9999 variants from this distribution, with 0.05<*p*
_{v}<0.99 and 0.05<*μ*
_{v}<5.

In each case, the population size is 10 million, with a starting wild-type frequency of 1%. Starting counts for each variant are simulated using a log-normal distribution of variant counts in the input time point such that the mean variant input count is 990 and the standard deviation of the distribution is 0.4 [16, 47]. Starting counts are independently generated for each replicate.

For each replicate, the starting population undergoes five rounds of selection. The count of each variant after binding (*k*
_{v,t}) is generated using a binomial distribution with parameters *n*=*c*
_{v,t−1}
^{'} and *p*=*p*
_{v}. The count of each variant after growth (*g*
_{v,t}) is generated using a negative binomial distribution with parameters *r*=*c*
_{v,t−1}
^{'} and *p* = *e*^{−μvΔt}, $\mathrm{\Delta}t=\frac{\delta \mathrm{ln}2}{{\mu}_{wt}}$. For these simulations, *δ*=2. The population count for each variant (*c*
_{v,t}
^{'}) is obtained by performing weighted random sampling with replacement, where the weight for each variant is proportional to *k*
_{v,t} or *g*
_{v,t} and the total population size was 10 million.

Read counts for each variant (*c*
_{v,t}) are simulated by performing weighted random sampling with replacement, where the weight for each variant is proportional to the population counts (*c*
_{v,t}
^{'}) and the average sequencing depth is 200 reads per variant (approximately 2 million reads per time point).

We simulate replicate noise by drawing a new variant effect from the variant effect distribution for 10% of variants (not including wild-type). These noisy variants were randomly chosen. This new variant effect was used to simulate one replicate and the other four replicates used the original effect. Noisy effects were split uniformly between the five replicates, such that 2% of the variants in each replicate were affected.

We simulate time point amplification and depletion noise by multiplying or dividing *c*
_{v,t}
^{'} by 50 before performing the sampling step to obtain *c*
_{v,t}. We randomly choose 10% of variants to be affected by noise, 5% subject to amplification and 5% subject to depletion, split uniformly among the five replicates. For each noisy variant in the chosen replicate, one time point (including input) was randomly chosen for amplification or depletion.

Python code for generating these simulated datasets is available as simdms v0.1 (DOI: 10.5281/zenodo.546311).

A region proximal to both lipid binding sites of the C2 domain of Phospholipase A2 (PLA2) was targeted for deep mutational scanning. Positions 94–97 of the C2 domain of mouse PLA2-alpha (ANYV) were fully randomized using a doped synthetic oligonucleotide. The library of C2 subdomains containing mutations was cloned into the AvrII and PpuMI sites of wild-type C2 domain in pGEM. The library was subcloned into phage arms and expressed on the surface of bacteriophage using the T7 phage display system according to the manufacturer’s instructions (Novagen T7Select 10-3b). The library was amplified in BLT5403 *E. coli* and variants were selected for their ability to bind to a lipid mixture containing ceramide 1-phosphate (C1P) [48]. The mouse PLA2-alpha cDNA was a generous gift from Michael Gelb, University of Washington. NiSepaharose Excel, capacity 10 mg/mL, was purchased from GE. Other reagents were purchased from Thermo-Fisher.

To select for C1P binding, lipid nanodiscs were developed as a bilayer affinity matrix. The His6-tagged membrane scaffold protein MSP1D1 [49] was expressed in BL21 *E. coli* from a pET28a plasmid and purified on nickel resin, then used to generate lipid nanodiscs comprising 30 mol% phosphatidylcholine, 20 mol% phosphatidylserine, 40 mol% phosphatidylethanolamine, and 10 mol% C1P [50]. To separate nanodiscs from large lipid aggregates and free protein, the mixture was subjected to gel filtration using a Superose 6 10/300 GL column (Pharmacia) and the major peak following the void volume was collected. To generate the affinity resin, 70 μg of nanodiscs (quantified by protein content) was incubated overnight at 4 °C with 10 μL nickel resin in 20 mM Tris pH 7.5 and 100 mM NaCl. The resin was washed twice in the same solution and used in phage binding reactions.

Phage expressing the C2 domain variant library were titered and diluted to a concentration of 5×10^{9} pfu/mL in 20 mM Tris pH 7.5 and 100 mM NaCl, then incubated with lipid nanodisc affinity resin plus 10 μM calcium in a final volume of 350 μL. After a 2-hour incubation at 4 °C, the resin was washed four times in 1 mL of the incubation buffer containing 20 mM imidazole. Phage bound to nanodiscs were eluted with 20 mM Tris pH 7.5 containing 500 mM imidazole. Phage from the elution were titered, amplified, and subjected to additional rounds of selection. Three replicate selections were performed on different days using the same input phage library.

Sequencing libraries were prepared by PCR amplifying the variable region using primers that append Illumina cluster generating and index sequences (Additional file 7) before sequencing using the Illumina NextSeq platform with a NextSeq high output kit (75 cycles, FC-404-1005). Reads were demultiplexed using bcl2fastq v2.17 (Illumina) with the arguments *bcl2fastq --with-failed-reads --create-fastq-for-index-reads --no-lane-splitting --minimum-trimmed-read-length 0 --mask-short-adapter-reads 0*. Quality was assessed using FastQC v0.11.3 [51]. Demultiplexed reads are available in the NCBI Sequence Read Archive, BioProject Accession PRJNA344387.

Raw reads were demultiplexed using a custom script based on three-nucleotide barcodes provided by the original authors [30]. The reads were analyzed in Enrich2 v1.0.0 as ten experimental conditions: five non-overlapping 30-base regions of the neuraminidase gene in either the presence or absence of oseltamivir. Reads were required to have a minimum quality score of 23 at all positions and contain no Ns. The five mutagenized regions were scored independently and then merged to create a single set of variant scores for each treatment. To be consistent with the original study, we removed variants containing multiple nucleotide changes with the exception of p.Ile117Ter and p.Thr226Trp that were individually validated. The *p* values for comparing variant scores to wild-type in each treatment and comparing variant scores between treatments were calculated using a *z*-test. All three sets of *p* values were jointly corrected for multiple testing using the qvalue package in R [52], and variants with a *q* value of less than 0.05 were reported as significant.

For previously published datasets, raw sequence files in FASTQ format were obtained from the respective authors. Datasets (Table 1) were analyzed independently using Enrich2 v1.0.0. The BRCA1 dataset was analyzed in a single run with separate experimental conditions for the yeast two-hybrid and phage display assays. For all datasets except neuraminidase, reads were required to have a minimum quality score of 20 at all positions and contain no Ns.

For the WW domain sequence function map (Fig. 1), scores and standard errors were calculated using weighted least squares linear regression in two technical replicates and the replicates were combined using the random-effects model as described.

Enrich2 is implemented in Python 2.7 and requires common dependencies for scientific Python. The graphical user interface is implemented using Tkinter. A deep mutational scanning experiment is represented as a tree of objects with four levels: experiment; condition; selection; and sequencing library. Each object’s data and metadata are stored in a single HDF5 file, including intermediate values calculated during analysis.

Enrich2 is designed to be run locally on a laptop computer and does not require a high-performance computing environment. Most analyses can be run overnight (Table 1). Run times in Table 1 were measured using a MacBook Pro Retina with 2.8 GHz Intel Core i7 processor and 16GB of RAM.

Additional file 4:^{(9.6K, xlsx)}

Individually validated variants of the neuraminidase gene. (XLSX 9 kb)

Additional file 5:^{(12K, xlsx)}

Variants with higher scores than wild-type in the presence of oseltamivir. (XLSX 11 kb)

We thank Stanley Fields for his support, advice, and comments. We thank Sara Rubin for editing the manuscript. We thank Lea Starita, Vanessa Gray, and Matt Rich for beta testing. We thank Daniel Esposito for his contributions to Enrich2 v1.1.0. We thank Bernie Pope and Matthew Wakefield for their software engineering and algorithm advice. We thank Galen Flynn for his help with the lipid nanodisc purification.

This work was supported by the National Institute of General Medical Sciences (1R01GM109110 and 5R24GM115277 to DMF and P41GM103533 to Stanley Fields); the National Institute of Biomedical Imaging and Bioengineering (5R21EB020277 to SMB and DMF); the Washington Research Foundation (Washington Research Foundation Innovation Postdoctoral Fellowship to HG); and the National Health and Medical Research Council of Australia (Program Grant 1054618 to ATP and TPS).

• Project name: Enrich2 (v1.1.0a)

• Project home page: https://github.com/FowlerLab/Enrich2

• Example dataset home page: https://github.com/FowlerLab/Enrich2-Example

• Documentation home page: http://enrich2.readthedocs.io/

• Archived version: 10.5281/zenodo.802188

• Operating systems: Platform independent

• Programming language: Python

• Other requirements: Python 2.7, multiple Python packages

• License: GNU GPLv3

• Any restrictions to use by non-academics: None

• Dataset accession numbers:

° Neuraminidase: BioProject PRJNA272490

° WW domain: SRA SRP002725

° C2 domain: BioProject PRJNA344387

Authors’ contributions

AFR, TPS, and DMF developed the statistical methods. AFR wrote the Enrich2 software and performed the data analysis. HG and AFR developed the simulation framework and generated the simulated datasets. NL, SMB, and DMF designed and performed the C2 domain deep mutational scan. ATP reviewed the Enrich2 codebase. AFR and DMF wrote the paper. All authors read and approved the final manuscript.

Ethics approval was not needed for this work.

The authors declare that they have no competing interests.

Springer Nature remains neutral with regard to jurisdictional claims in published maps and institutional affiliations.

**Electronic supplementary material**

The online version of this article (doi:10.1186/s13059-017-1272-5) contains supplementary material, which is available to authorized users.

1. Fowler DM, Araya CL, Fleishman SJ, Kellogg EH, Stephany JJ, Baker D, et al. High-resolution mapping of protein sequence-function relationships. Nat Methods. 2010;7:741–6. doi: 10.1038/nmeth.1492. [PMC free article] [PubMed] [Cross Ref]

2. Fowler DM, Fields S. Deep mutational scanning: a new style of protein science. Nat Methods. 2014;11:801–7. doi: 10.1038/nmeth.3027. [PMC free article] [PubMed] [Cross Ref]

3. Majithia AR, Tsuda B, Agostini M, Gnanapradeepan K, Rice R, Peloso G, et al. Prospective functional classification of all possible missense variants in PPARG. Nat Genet. 2016;48:1570–5. doi: 10.1038/ng.3700. [PMC free article] [PubMed] [Cross Ref]

4. Starita LM, Young DL, Islam M, Kitzman JO, Gullingsrud J, Hause RJ, et al. Massively Parallel Functional Analysis of BRCA1 RING Domain Variants. Genetics. 2015;200:413–22. doi: 10.1534/genetics.115.175802. [PubMed] [Cross Ref]

5. Bank C, Hietpas RT, Jensen JD, Bolon DNA. A systematic survey of an intragenic epistatic landscape. Mol Biol Evol. 2015;32:229–38. doi: 10.1093/molbev/msu301. [PMC free article] [PubMed] [Cross Ref]

6. Podgornaia AI, Laub MT. Protein evolution. Pervasive degeneracy and epistasis in a protein-protein interface. Science. 2015;347:673–7. doi: 10.1126/science.1257360. [PubMed] [Cross Ref]

7. Rockah-Shmuel L, Tóth-Petróczy Á, Tawfik DS. Systematic mapping of protein mutational space by prolonged drift reveals the deleterious effects of seemingly neutral mutations. PLoS Comput Biol. 2015;11:e1004421. doi: 10.1371/journal.pcbi.1004421. [PMC free article] [PubMed] [Cross Ref]

8. Stiffler MA, Hekstra DR, Ranganathan R. Evolvability as a function of purifying selection in TEM-1 β-Lactamase. Cell. 2015;160:882–92. doi: 10.1016/j.cell.2015.01.035. [PubMed] [Cross Ref]

9. Wu NC, Dai L, Olson CA, Lloyd-Smith JO, Sun R. Adaptation in protein fitness landscapes is facilitated by indirect paths. Elife. 2016;5:e16965. [PMC free article] [PubMed]

10. Adkar BV, Tripathi A, Sahoo A, Bajaj K, Goswami D, Chakrabarti P, et al. Protein model discrimination using mutational sensitivity derived from deep sequencing. Structure. 2012;20:371–81. doi: 10.1016/j.str.2011.11.021. [PubMed] [Cross Ref]

11. Sahoo A, Khare S, Devanarayanan S, Jain PC, Varadarajan R. Residue proximity information and protein model discrimination using saturation-suppressor mutagenesis. Elife. 2015;4:e09532. doi: 10.7554/eLife.09532. [PMC free article] [PubMed] [Cross Ref]

12. Fowler DM, Stephany JJ, Fields S. Measuring the activity of protein variants on a large scale using deep mutational scanning. Nat Protoc. 2014;9:2267–84. doi: 10.1038/nprot.2014.153. [PMC free article] [PubMed] [Cross Ref]

13. Hiatt JB, Patwardhan RP, Turner EH, Lee C, Shendure J. Parallel, tag-directed assembly of locally derived short sequence reads. Nat Methods. 2010;7:119–22. doi: 10.1038/nmeth.1416. [PMC free article] [PubMed] [Cross Ref]

14. Starita LM, Pruneda JN, Lo RS, Fowler DM, Kim HJ, Hiatt JB, et al. Activity-enhancing mutations in an E3 ubiquitin ligase identified by high-throughput mutagenesis. Proc Natl Acad Sci U S A. 2013;110:E1263–72. doi: 10.1073/pnas.1303309110. [PubMed] [Cross Ref]

15. Bloom JD. An experimentally determined evolutionary model dramatically improves phylogenetic fit. Mol Biol Evol. 2014;31:1956–78. doi: 10.1093/molbev/msu173. [PMC free article] [PubMed] [Cross Ref]

16. Matuszewski S, Hildebrandt ME, Ghenu A-H, Jensen JD, Bank C. A statistical guide to the design of deep mutational scanning experiments. Genetics. 2016;204:77–87. doi: 10.1534/genetics.116.190462. [PubMed] [Cross Ref]

17. Starita LM, Fields S. Deep mutational scanning: a highly parallel method to measure the effects of mutation on protein function. Cold Spring Harb Protoc. 2015;2015:711–4. [PubMed]

18. Kowalsky CA, Klesmith JR, Stapleton JA, Kelly V, Reichkitzer N, Whitehead TA. High-resolution sequence-function mapping of full-length proteins. PLoS One. 2015;10:e0118193. doi: 10.1371/journal.pone.0118193. [PMC free article] [PubMed] [Cross Ref]

19. Fowler DM, Araya CL, Gerard W, Fields S. Enrich: software for analysis of protein function by enrichment and depletion of variants. Bioinformatics. 2011;27:3430–1. doi: 10.1093/bioinformatics/btr577. [PMC free article] [PubMed] [Cross Ref]

20. Hietpas RT, Jensen JD, Bolon DNA. Experimental illumination of a fitness landscape. Proc Natl Acad Sci U S A. 2011;108:7896–901. doi: 10.1073/pnas.1016024108. [PubMed] [Cross Ref]

21. Patwardhan RP, Lee C, Litvin O, Young DL, Pe’er D, Shendure J. High-resolution analysis of DNA regulatory elements by synthetic saturation mutagenesis. Nat Biotechnol. 2009;27:1173–5. doi: 10.1038/nbt.1589. [PMC free article] [PubMed] [Cross Ref]

22. Araya CL, Fowler DM, Chen W, Muniez I, Kelly JW, Fields S. A fundamental protein property, thermodynamic stability, revealed solely from large-scale measurements of protein function. Proc Natl Acad Sci U S A. 2012;109:16858–63. doi: 10.1073/pnas.1209751109. [PubMed] [Cross Ref]

23. Rich MS, Payen C, Rubin AF, Ong GT, Sanchez MR, Yachie N, et al. Comprehensive analysis of the SUL1 promoter of Saccharomyces cerevisiae. Genetics. 2016;203:191–202. doi: 10.1534/genetics.116.188037. [PubMed] [Cross Ref]

24. Melnikov A, Rogov P, Wang L, Gnirke A, Mikkelsen TS. Comprehensive mutational scanning of a kinase in vivo reveals substrate-dependent fitness landscapes. Nucleic Acids Res. 2014;42:e112. doi: 10.1093/nar/gku511. [PMC free article] [PubMed] [Cross Ref]

25. Roscoe BP, Thayer KM, Zeldovich KB, Fushman D, Bolon DNA. Analyses of the effects of all ubiquitin point mutants on yeast growth rate. J Mol Biol. 2013;425:1363–77. doi: 10.1016/j.jmb.2013.01.032. [PMC free article] [PubMed] [Cross Ref]

26. Jiang L, Mishra P, Hietpas RT, Zeldovich KB, Bolon DNA. Latent effects of Hsp90 mutants revealed at reduced expression levels. PLoS Genet. 2013;9:e1003600. doi: 10.1371/journal.pgen.1003600. [PMC free article] [PubMed] [Cross Ref]

27. Mavor D, Barlow K, Thompson S, Barad BA, Bonny AR, Cario CL, et al. Determination of ubiquitin fitness landscapes under different chemical stresses in a classroom setting. Elife. 2016;5:e15802. doi: 10.7554/eLife.15802. [PMC free article] [PubMed] [Cross Ref]

28. Bank C, Hietpas RT, Wong A, Bolon DN, Jensen JD. A Bayesian MCMC approach to assess the complete distribution of fitness effects of new mutations: uncovering the potential for adaptive walks in challenging environments. Genetics. 2014;196:841–52. doi: 10.1534/genetics.113.156190. [PMC free article] [PubMed] [Cross Ref]

29. Borenstein M, Hedges LV, Higgins JPT, Rothstein HR. Introduction to meta-analysis. Chichester: Wiley; 2009.

30. Jiang L, Liu P, Bank C, Renzette N, Prachanronarong K, Yilmaz LS, et al. A balance between inhibitor binding and substrate processing confers influenza drug resistance. J Mol Biol. 2016;428:538–53. doi: 10.1016/j.jmb.2015.11.027. [PubMed] [Cross Ref]

31. Forsyth CM, Juan V, Akamatsu Y, DuBridge RB, Doan M, Ivanov AV, et al. Deep mutational scanning of an antibody against epidermal growth factor receptor using mammalian cell display and massively parallel pyrosequencing. MAbs. 2013;5:523–32. doi: 10.4161/mabs.24979. [PMC free article] [PubMed] [Cross Ref]

32. Kim I, Miller CR, Young DL, Fields S. High-throughput analysis of in vivo protein stability. Mol Cell Proteomics. 2013;12:3370–8. doi: 10.1074/mcp.O113.031708. [PMC free article] [PubMed] [Cross Ref]

33. Kosuri S, Goodman DB, Cambray G, Mutalik VK, Gao Y, Arkin AP, et al. Composability of regulatory sequences controlling transcription and translation in Escherichia coli. Proc Natl Acad Sci U S A. 2013;110:14024–9. doi: 10.1073/pnas.1301301110. [PubMed] [Cross Ref]

34. Melamed D, Young DL, Gamble CE, Miller CR, Fields S. Deep mutational scanning of an RRM domain of the Saccharomyces cerevisiae poly(A)-binding protein. RNA. 2013;19:1537–51. doi: 10.1261/rna.040709.113. [PubMed] [Cross Ref]

35. Tinberg CE, Khare SD, Dou J, Doyle L, Nelson JW, Schena A, et al. Computational design of ligand-binding proteins with high affinity and selectivity. Nature. 2013;501:212–6. doi: 10.1038/nature12443. [PMC free article] [PubMed] [Cross Ref]

36. Guy MP, Young DL, Payea MJ, Zhang X, Kon Y, Dean KM, et al. Identification of the determinants of tRNA function and susceptibility to rapid tRNA decay by high-throughput in vivo analysis. Gene Dev. 2014;28:1721–32. doi: 10.1101/gad.245936.114. [PubMed] [Cross Ref]

37. Reich LL, Dutta S, Keating AE. SORTCERY-A high-throughput method to affinity rank peptide ligands. J Mol Biol. 2015;427:2135–50. doi: 10.1016/j.jmb.2014.09.025. [PMC free article] [PubMed] [Cross Ref]

38. Levine HA, Nilsen-Hamilton M. A mathematical analysis of SELEX. Comput Biol Chem. 2007;31:11–35. doi: 10.1016/j.compbiolchem.2006.10.002. [PMC free article] [PubMed] [Cross Ref]

39. Levitan B. Stochastic modeling and optimization of phage display. J Mol Biol. 1998;277:893–916. doi: 10.1006/jmbi.1997.1555. [PubMed] [Cross Ref]

40. Levin AM, Weiss GA. Optimizing the affinity and specificity of proteins with molecular display. Mol Biosyst. 2006;2:49–57. doi: 10.1039/B511782H. [PubMed] [Cross Ref]

41. Brockmann E-C. Selection of stable scFv antibodies by phage display. Methods Mol Biol. 2012;907:123–44. doi: 10.1007/978-1-61779-974-7_7. [PubMed] [Cross Ref]

42. Needleman SB, Wunsch CD. A general method applicable to the search for similarities in the amino acid sequence of two proteins. J Mol Biol. 1970;48:443–53. doi: 10.1016/0022-2836(70)90057-4. [PubMed] [Cross Ref]

43. Plackett RL. The analysis of categorical data. 2. New York: MacMillan; 1981.

44. Seber GAF. Linear regression analysis. New York: Wiley; 1977.

45. Demidenko E. Mixed models: theory and applications with R. 2. Hoboken: Wiley; 2013.

46. Spill F, Weinstein ZB, Irani Shemirani A, Ho N, Desai D, Zaman MH. Controlling uncertainty in aptamer selection. Proc Natl Acad Sci U S A. 2016;113:12076–81. doi: 10.1073/pnas.1605086113. [PubMed] [Cross Ref]

47. Wrenbeck EE, Klesmith JR, Stapleton JA, Adeniran A, Tyo KEJ, Whitehead TA. Plasmid-based one-pot saturation mutagenesis. Nat Methods. 2016;13:928–30. doi: 10.1038/nmeth.4029. [PubMed] [Cross Ref]

48. Lamour NF, Subramanian P, Wijesinghe DS, Stahelin RV, Bonventre JV, Chalfant CE. Ceramide 1-phosphate is required for the translocation of group IVA cytosolic phospholipase A2 and prostaglandin synthesis. J Biol Chem. 2009;284:26897–907. doi: 10.1074/jbc.M109.001677. [PMC free article] [PubMed] [Cross Ref]

49. Dalal K, Chan CS, Sligar SG, Duong F. Two copies of the SecY channel and acidic lipids are necessary to activate the SecA translocation ATPase. Proc Natl Acad Sci U S A. 2012;109:4104–9. doi: 10.1073/pnas.1117783109. [PubMed] [Cross Ref]

50. Denisov IG, Grinkova YV, Lazarides AA, Sligar SG. Directed self-assembly of monodisperse phospholipid bilayer Nanodiscs with controlled size. J Am Chem Soc. 2004;126:3477–87. doi: 10.1021/ja0393574. [PubMed] [Cross Ref]

51. Andrews S. FastQC A quality control tool for high throughput sequence data. http://www.bioinformatics.babraham.ac.uk/projects/fastqc/. Accessed 8 Sept 2016.

52. Storey JD. A direct approach to false discovery rates. J Roy Stat Soc B Wiley Online Library. 2002;64:479–98. doi: 10.1111/1467-9868.00346. [Cross Ref]

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